Explaining Electoral Support for the Radical Right

 

1 Introduction: Voting for the Radical Right

Within the larger field of Radical Right studies, the question of why people vote for Radical Right Parties (RRPs) has attracted a large (perhaps disproportionally so) chunk of scholarly attention. There are at least three reasons for this. First, the early (and rather humble) electoral successes of the Radical Right in Western Europe during the early 1980s stirred memories of the 1920 and 1930s, when parties such as the Italian Fascists or the German Nazis rose from obscurity to overturn democracy (Prowe, 1994). Given these traumatic experiences, scholars were understandably eager to analyse the motives behind such potentially fatal electoral choices.

Second, when it became increasingly clear that the most electorally successful of these RRPs were not just clones of the old fascist right of the inter war years but rather belonged to a new party family (Mudde, 1996), researchers wanted to understand the social forces that brought about the rise of this largely unexpected phenomenon. After all, even non-extremist RRPs are still widely seen as problematic, because they promote a political ideal that has been dubbed “illiberal democracy” (Mudde, 2007), and often disrupt the political process.

Third, support for the Radical Right displays an unusual degree of variation across time and space. In Southern Europe, Cyprus (until 2016), Malta, Portugal and Spain never had a relevant RRP, whereas RRPs have been more or less consistently successful in Austria, Denmark, France, Italy, and Norway. Electoral support for the Radical Right has been volatile in Germany, Greece, Sweden and the UK. In the Netherlands, which featured extremist but tiny right-wing parties in the 1980s and 1990s, modern RRPs only emerged in the early 2000s. As of 2016, the radical right PVV is the country’s largest political party in terms of voting intentions. Belgium provides perhaps the most striking example of variability: While the Walloon National Front always remained at the margins in Wallonia, the Vlaams Blok/Vlaams Belang went from strength to strength in the Flemish part of the country during the 1990s and early 2000s, but lost roughly three quarters of their support between 2004 and 2014. To summarise, there is ample reason for treating support for the Radical Right as an unusual and potentially even dangerous phenomenon.

The most obvious way to study Radical Right voting would be to apply the standard tools of electoral research. Modern election studies usually rely on an eclectic blend of variables and alleged mechanisms, but at the core, there is usually the assumption that voters respond to short-term factors (candidates and political issues) on the one hand, and long-to-medium forces (party loyalties, value orientations, ideological convictions and group memberships) on the other. Almost sixty years ago, Angus Campbell and his associates (Campbell, 1960) have proposed a conceptual framework that encompasses these and other variables: In their “funnel of causality” metaphor, the proximate determinants of a given electoral choice are causally linked to more distant antecedents, forming a “funnel” that gets wider as more and more stable attitudes and earlier events are being considered. Decades of criticism not withstanding, this framework still explicitly or implicitly undergirds most empirical research into voting behaviour.

In the subfield of Radical Right voting, however, researchers habitually seem to ignore most of what constitutes the “normal science” (Kuhn, 1962) of electoral research, either because they are unaware of it, or because they are chiefly interested in “deeper” explanations that are located towards the far side of the funnel. Nonetheless, the funnel metaphor still provides a useful template for organising and comparing competing and complementary explanations for Radical Right electoral support.

However, the distinction between “supply side” and “demand side” factors, which can be traced back to an early article by Klaus von Beyme (Beyme, 1988), proved to be a much more popular schema for structuring potential explanations. Unfortunately, it is not entirely clear what is meant by “supply” and “demand” in this context and whether these two exhaust the full set of relevant factors, although the dichotomy has a certain heuristic value: The notion of a “supply side” usually refers to all variables pertaining to the RRP itself. This includes, but is not limited to, the stylistic and substantive content of the party manifesto and other texts, speeches or statements produced by the party, the party’s organisational structure and resources, and the presence or absence of a “charismatic leader”. The “demand side”, on the other hand, encompasses traits, experiences and attitudes that may predispose voters to support an RRP.

A number of other relevant factors, however, do not sit easily within the confines of this dichotomy. The ideological positions of mainstream right parties, for instance, could be considered part of the “supply” in a wider sense, but the same is not true for institutional variables such as the electoral system or the degree of decentralisation. These features of the wider political system may explain why would-be political entrepreneurs decide to enter the political arena to provide a RRP supply, or why a given demand for RRP policies may help or hurt the mainstream right parties. Put differently, many institutional factors should be seen as mediators of supply and demand rather than as members of either category. Other system-level variables – most prominently unemployment and immigration – are best understood as distal causes of demand, or as an incentives to provide supply.

Therefore, it seems more fruitful to distinguish between variables on the micro, meso, and macro level, and the remainder of this chapter will proceed accordingly. Most approaches, however, more or less explicitly follow the logic of a multi-level explanation (Coleman, 1994), requiring occasional cross-references between the sections.

The literature on this topic is already vast and keeps on growing quickly. My self-consciously eclectic bibliography on the Radical Right in Europe (http://www.kai-_arzheimer.com/extreme-_right-_western-_europe-_bibliography), which is nowhere near complete, currently stands at more than 600 titles. The literature review in this chapter is therefore by necessity highly selective and idiosyncratic: I will focus on (Western) Europe, and on a small number of contributions that I consider landmarks. Although comparative multi-level analyses are now something like the gold standard in the field, I will also consider single-country case studies where they present results that (probably) generalise beyond the polity in question, or designs that are of a more general interest. Moreover, while there is always the danger of aggregation bias lurking in the background, I will frequently discuss findings from field-defining aggregate studies, without re-iterating the usual warnings about the ecological fallacy (Robinson, 1950) time and again. Consider yourself trigger-warned.

2 Micro-level Factors

2.1 Party Identification

Party identification is arguably the most important factor when it comes to explaining voting decisions, but it is conspicuously underrepresented in the literature on the Radical Right. One possible explanation for this is the fact that party identification is supposed to be acquired through years, if not decades of political socialisation. As many RRPs only rose to prominence in the 1980s and 1990s, identification with them could hardly be a major factor behind their ascendancy. A a consequence, most early studies completely ignored party identification, and one of those few assessing its effect (based on data from the mid-1990s) concluded that “the identification motive is clearly significantly under-represented among VB [Vlaams Blok] voters” (Swyngedouw, 2001, p. 228).

A more modern approach highlights the negative effect of identifications with other parties. Building on the notion (derived from the older literature, e.g. Kitschelt (1995) and Ignazi (2003)) that the rise of the Radical Right only became possible once there was a sufficiently large pool of voters that were no longer attached to any of the established parties, Arzheimer and Carter (2009a) focus on (the lack of) identifications with mainstream right-wing parties. Using data from the 2002/03 wave of the European Social Survey, they demonstrate that voters who are still attached to a Christian Democratic or Conservative party almost never vote for a Radical Right party. Put differently, they see the absence of other identifications as a necessary (if insufficient) pre-condition for Radical Right-wing voting. However, some of the most successful RRPs (e.g. the French National Front, the Austrian Freedom Party, the Danish People’s Party or the Norwegian Progress Party) have been electorally relevant for two decades or more now, so the impact of identifying with the RRP should be modeled, too, but very few authors (e.g. Arzheimer, 2009b) account for this potential positive effect of party identification.

2.2 Candidates: The (ir)relevance of charismatic leaders

While party identifications have been more or less neglected as a key explanatory variable for RRP support, candidates and more specifically “charismatic” party leaders have attracted a great deal of attention (e.g. Taggart, 1995). There are two reasons for this: First, many observers mistook the rise of the RRPs in the 1980s for a “Return of the Führers” of the 1920s (Prowe, 1994). Second, many RRPs appeared to be personal parties, especially during the break-through phase (Eatwell, 2005, p. 106). Third, agency is always more attractive than structure.

However, what is meant by “charisma” is not usually clear. There are serious doubts that Weberian “charisma” – a personal bond between the (party) leader and his followers – was in any way relevant for the rise of the Radical Right (Eatwell, 2005), and even those two parties most commonly associated with their “charismatic” leaders – Joerg Haider’s Austrian Freedom Party and Jean-Marie Le Pen’s French National Front – underwent a process of “institutionalisation” (Pedahzur and Brichta, 2002). Even more importantly for the question of electoral behaviour, Brug and Mughan (2007) demonstrate that RRPs benefit from candidate effects in exactly the same way as established parties: While having an appealing candidate is certainly linked to greater electoral support, the magnitude of this effect is not larger than it is for other parties.

2.3 Issues, Ideology and Value Orientations

2.3.1 Pure Protest Voting, Anti-Immigrant Sentiment, and Unemployment (threat)

When it comes to explaining Radical Right support, the notion of a “pure protest vote” is still prominent. In its most extreme guise, the pure protest thesis claims that Radical Right support is driven by feelings of alienation from the political elites and the political system that are completely unrelated to policies or values and hence have nothing to do with the Radical Right’s political agenda (Eatwell, 2000). A more realistic variety of the protest thesis suggests that voters do indeed care about policies but hold less extreme preferences than the Radical Right manifestos would suggest. In this scenario, voters instrumentally support the Radical Right in the hope that mainstream right parties will reconsider their position and move somewhat closer to the Radical Right without copying all of their policies. Once the mainstream right has made this adjustment, Radical Right support would collapse. This logic is akin to directional voting (Merrill and Grofman, 1999) but puts more emphasis on emotions.

Empirically, pure protest voting remains elusive. Starting with Billiet and Witte’s (1995) study of Vlaams Blok support in the 1991 General Election in Belgium, a host of single-country and comparative studies have demonstrated time and again that anti-immigrant sentiment is the single most important driver of the Radical Right vote (Mayer and Perrineau, 1992; Brug, Fennema, and Tillie, 2000; Brug and Fennema, 2003; Norris, 2005; Mughan and Paxton, 2006; Arzheimer, 2009b; Ford, Goodwin, and Cutts, 2011). That does not mean that the prototypical voter of the Radical Right is not alienated from the political elites and susceptible to the populist rhetoric of many RRPs. But the vast majority of their voters support the Radical Right because of their anti-immigrant claims and demands, and their sense of frustration and distrust may very well result from their political preferences on immigration not being heeded by the mainstream parties.

Anti-immigrant sentiment is a handy but slightly awkward catch-all term for negative attitudes towards immigrants, immigration, and immigration policies. In a seminal contribution, Rydgren (2008) distinguishes between “immigration sceptics”, “xenophobes”, and “racists”. For Rydgren (2008, pp. 741-744), xenophobes have a latent disposition to react with fear and aversion to outsiders, but this only becomes an issue if the number of outsiders is too high by some subjective standard, or if the outsiders otherwise seem to pose a threat to in-group. Racists always hold outsiders in contempt irrespective of any exposure to “strangers”, with “classic” racism being based on notions of biological hierarchies, whereas “modern” or “cultural” racism subscribes to the idea of incompatible but (nominally) coequal cultures.1 Finally, immigration sceptics want to reduce the number of immigrants in their native country (Rydgren, 2008, p. 738), but not necessarily because they hold racist or xenophobic attitudes. As Rydgren (2008, p. 740) suggests, the most plausible structure for these attitudes is a nested one, where xenophobes form a subgroup of the immigration sceptics and racists form a subgroup of the xenophobes.

The distinction between immigration sceptics, xenophobes, and racists is particularly useful because not all Radical Right voters are full-blown racists. Moreover, many of the approaches that are discussed in the literature may help to explain deep-seated, stable racism but not necessarily a more specific and volatile scepticism regarding current immigration policies.

“Deep” explanations for Radical Right support have been developed since at least the 1930s. The monographs and articles on the roots of rightist political views fill several libraries by now and any attempt to classify them is crude by necessity. Nonetheless, it makes sense to distinguish between three very broad groups.

A first class of explanations focuses on personality traits2, with authoritarianism being the most prominent amongst them. Authoritarianism as a concept is most closely associated with the (controversial) Berkeley Study (Adorno et al., 1950) but has more recently been modernised and promoted by Bob Altemeyer (1981; 1996). For Altemeyer, Right-Wing Authoritarianism (RWA) consists of three key elements: a desire to submit to established and legitimate authorities (authoritarian submission), a hostility towards deviants and other out-groups (authoritarian aggression), and an exaggerated respect for traditions and social norms (conventionalism).

Authoritarianism and similar concepts such as dogmatism (Rokeach, 1960) or tough-mindedness (Eysenck, 1954) go a long way towards explaining the relevance of xenophobia and the appeal of other right-wing ideas and movements to some voters, but there are a few important caveats. First, compared to classic right-wing extremist groups, authoritarianism is much less important for the ideology of the modern populist Radical Right (Mudde, 2007). Unlike the Fascists or the Nazis of the interwar period, the most successful of these parties do not seek to replace democracy by some authoritarian type of regime but rather promote a narrow, “illiberal” concept of democracy. Second, support for the Radical Right has surged (and sometimes declined) over relatively short periods, whereas personality traits are by definition stable. They may thus help us to explain why there is potential for authoritarian parties in the first place. The exploitation of this potential by political entrepreneurs and the channeling of this general hostility towards out-groups into a more specific anti-immigrant sentiment, however, are political processes that must be understood by means of different concepts.

Theories of group conflict and deprivation form a second and more immediately relevant cluster of explanations. This cluster can be subdivided in four broad categories

  1. Theories of “realistic group conflict” (RGCT) and “ethnic competition” (EC)
  2. Theories of “status politics” and “symbolic racism”
  3. Theories of “social identity”
  4. Theories of “scapegoating”

The ordering is deliberate: From the top to the bottom, these approaches put less and less emphasis on material conflicts and conscious mental processes and instead focus on the importance of visceral hostility (which might still be induced by political entrepreneurs) towards members of the out-group.

Both for proponents of RGCT (see Jackson, 1993 for a review) and EC (e.g. Bélanger and Pinard, 1991), tensions between (ethnic) groups are rooted in conflicts over the distribution of material resources in a society, which is often perceived as unfair. The main difference between both approaches is that RGCT is more interested in the micro-dynamics of group psychology whereas EC is primarily concerned with the societal level. Either way, the distributional conflict is couched in collective terms, even if the resource in question is a personal good (e.g. a secure job). Both strands of the literature as well as the other approaches discussed in this section are therefore closely related to classic theories of collective relative deprivation (Runciman, 1966, pp. 33-34, see also Ellemers, 2002 and Taylor, 2002). While students of electoral behaviour rarely investigate the lengthy and complex causal chains that link social change, group dynamics, and inter-ethnic contacts to psychological processes, feelings of material threat that is allegedly posed by immigrants have become a staple explanatory variable for analysing anti-immigrant sentiment, and by implication the Radical Right’s electoral support. On the contextual level, (potential) exposure to material threats if often captured by incorporating macro-economic variables in statistical models of Radical Right voting (see below).

Similarly, proponents of the “status politics” approach (e.g. Hofstadter, 2002b) argue that (recent) immigrants are perceived as a collective threat by members of the in-group. Here, the collective good in question is not a material one but rather the collective social status of the in-group, or the cultural hegemony of their values, norms, and social practices (Hofstadter, 2002a) – ideas which in turn bear some resemblance with the idea of “symbolic racism” (Kinder and Sears, 1981; see Walker, 2001 for a critical review of this and some related concepts). Again, psephologists usually take the alleged causal mechanisms for granted and focus on the effect of perceived cultural threats on anti-immigrant sentiment and the Radical Right vote.

(Modern) theories of social identity provide another approach for explaining anti-immigrant sentiment. “Social Identity Theory” (SIT) and its successor, “Self-Categorisation Theory” (SCT), were developed in response to an empirical puzzle: Even in a “minimal effects” experimental setting where subjects were randomly assigned to socially meaningless groups, where there was no interaction whatsoever between subjects, and no material incentive to put members of the out-group at a disadvantage, a large proportion of subjects was willing to discriminate against the outsiders. Tajfel and Turner (1986) interpret this unexpected finding as the result of a cognitive process during which one’s social identity becomes the yardstick for assessing a given situation, whereas the importance of one’s personal identity declines. As a corrolary, members of the out-group are subject to a process of stereotyping. In combination with an innate desire for positive distinctiveness, stereotyping and self-stereotyping can bring about discrimination and prejudice against out-group members, because they represent one avenue towards a more positive self-image. However, whether discrimination actually occurs depends on a number of conditions (Reynolds and Turner, 2001, p. 166). Crucially, these mechanisms are independent of any material or cultural threat that the out-group may seem to pose to the members of the in-group.

Once more, psephologists have mostly ignored the details and instead focused on the impact of a single variable (identity) on Radical Right voting intentions, and even this alleged mechanism is often problematic, because most items available in representative surveys do not capture the complexity of the concept. Nonetheless, SIT/SCT has the potential to make a crucial contribution to a fuller explanation of the Radical Right vote: While most group dynamic processes must remain under the radar of mass surveys, SIT/SCT informs experimental and observational research on the conditions under which stereotypes and prejudices that may result in anti-immigrant sentiment become activated. It also provides a useful framework for the analysis of party documents and social and mass media content, which play an ever more important role in the study of Radical Right electoral support.

Finally, theories of “scapegoating” need to be addressed. These hark back to the late 1930s (Dollard et al., 1939) and have even older roots in the Sumner’s early work on ethnocentrism (Sumner, 1906), maintain that members of the ethnic majority who experience feelings of frustration and deprivation that are objectively unrelated to the presence of other ethnic groups nonetheless turn towards immigrants simply because those provide a conveniently defenceless target for the in-group members’ aggression. Due to the “cognitive turn” in social psychology, theories of scapegoating have somewhat fallen out of fashion, and for the applied psephologist relying on secondary data analysis, the result of simple scapegoating will often be indistinguishable from the more complex stereotyping processes.

All theories of group conflict are complemented by the “contact hypothesis”, which maintains that under certain favourable conditions, inter-ethnic contacts (which often presuppose immigration) can reduce prejudice (Pettigrew and Tropp, 2008) and hence anti-immigrant sentiment. Some of the newer research aims at incorporating the contact hypothesis by either using micro-level information on inter-ethnic contact or by deriving the probability of such contacts from small-area data on the spatial distribution of ethnic groups. Unfortunately, both approaches are subject to endogeneity bias, because voters who are less prejudiced are more likely to seek inter-ethnic contacts.

2.3.2 Anti Post-Materialism and Other Social Attitudes

A Silent Counter-Revolution? Immigration emerged as the core issue of the Radical Right in Western Europe and Australia in the mid-1980s, making anti-immigrant sentiment the single most important attitudinal driver of Radical Right support. In Central and Eastern Europe (CEE), hostility towards ethnic minorities seems to act as the functional equivalent. But very few RRPs have ever been single-issue parties (Mudde, 1999). Many of them have a broader right-wing agenda, and Radical Right support has been linked to a host of other attitudes than anti-immigrant sentiment.

The Rise of the RRP family in the 1980s and early 1990s has therefore been interpreted as a reaction to large-scale social change.3 In a seminal article, Ignazi (1992) claims that these new right-wing parties embody the backlash against post-materialism and the New Left politics which it has inspired: a “silent counter-revolution”. Similarly, Kitschelt (1995) has argued that globalisation has created a new class of authoritarian private-sector workers, who combine market-liberal preferences with an authoritarian outlook on society and find their political representation in the Radical Right. While the market-liberalism of the Radical Right’s electorate remains elusive (Kitschelt and McGann, 2003; Arzheimer, 2009b; Mayer, 2013), it has become ever more evident that non-traditional working-class voters form the Radical Right’s core electoral base (see the contributions in Rydgren, 2013).

Moral conservatism, homophobia and more generally anti-postmaterialism may have played a role, too (and probably are still relevant for party members and activists), but they seem to be much less important than they were for the classic Extreme Right, at least in some countries. As early as 1988, the French FN voters were slightly “more permissive in sexual matters” than the voters of the mainstream right (Mayer and Perrineau, 1992, p. 130). 25 years later, the FN is lead by a single mother of three, twice divorced (Mayer, 2013, p. 175), whose attendance at homophobic rallies seems to be more a matter of strategy than of convictions. Even more strikingly, the Lijst Pim Fortuyn, the Netherland’s first successful RRP, was founded and led by an openly gay libertine (Akkerman, 2005), and its de facto successor, the PVV, claims that defending the freedom of the LGBT community is part of their commitment to Dutch values. But even in the Netherlands, culturally progressive values are not an important driver of the RRP vote, at least not when anti-immigrant sentiment is controlled for (De Koster et al., 2014). One way or the other, for many RRP voters in Western Europe, homophobia and social conservatism do not seem to matter too much any more.

Religion The Extreme Right of the interwar years could be roughly divided in two groups (Camus, 2007): In some cases (most prominently Portugal and Spain), they aligned themselves with the most authoritarian and reactionary elements of the (Catholic) church. In other instances (e.g. Germany and Austria after the “Anschluss”), the Extreme Right distanced itself from Christianity and/or relied on the traditional loyalty of the (Protestant) church to the political leadership.

Today’s RRPs have inherited some of this historical baggage. While religious conservatism may inspire some of their members and voters (see the previous section), church leaders have often spoken out against the Radical Right’s anti-immigrant policies. To complicate matters further, the Radical Right is now often couching their anti-immigrant message in terms of a clash between “Western Values” and “Islam”. In a sense, criticising Islam abroad and at home has become the socially acceptable alternative to more openly xenophobic statements (Zúquete, 2008).

In a bid to disentangle this relationship, Arzheimer and Carter (2009a) estimate a Structural Equation Model of religiosity, anti-immigrant sentiment, party identification with mainstream right parties, and Radical Right voting intentions in seven West European countries. Their results show that in the early 2000s, religiosity had no significantly positive or negative effect on either anti-immigrant sentiment or RRP voting intentions. Religious people are, however, much more likely to identify with a mainstream right party, which in turn massively reduces the likelihood of an RRP vote. Using a slightly different model and data collected in 2008, Immerzeel, Jaspers, and Lubbers (2013) arrive at very similar conclusions.

Crime Law and order politics is traditionally the domain of both the mainstream and the Radical Right (Bale, 2003), with some authors going as far as saying that the Radical Right “owns” the crime issue (Smith, 2010). At any rate, talking about crime and immigration is a core frame of Radical Right discourses (Rydgren, 2008). Data from the European Social Survey clearly show that many West Europeans associate immigration with crime, and panel data from Germany suggest that that worries about crime have a substantial effect on anti-immigrant sentiment (Fitzgerald, Curtis, and Corliss, 2012). Many authors subsume such immigration-related crime fears into the larger complex of subjective threat that immigration poses to susceptible voters. Others model the effect of objective crime figures on the Radical Right vote (see below).

Euroscepticism Mudde (2007) has convincingly argued that nativism, i.e. the desire for an ethnically homogeneous nation state, forms the core of the Radical Right’s ideology. Accordingly, RRPs reject the European Union as a general rule, although Vasilopoulou (2011) has demonstrated that opposition to the European projects is by no means uniform within the Radical Right camp. Unsurprisingly, individual eurosceptic attitudes come up as predictors of Radical Right voting intentions in some studies (e.g. Arzheimer, 2009a; Brug, Fennema, and Tillie, 2005), although anti-immigrant and even general dissatisfaction with the elites exert a stronger effect (Werts, Scheepers, and Lubbers, 2013). Given that at least some countries feature leftist eurosceptic parties whose voters hold opinions which differ markedly from those of the RRP voters (Evans, 2000; Elsas and Brug, 2015), it seems safe to assume that euroscepticism per se does not predispose voters to support the Radical Right but needs to be linked to more general nativist beliefs.

3 Meso-level Factors

3.1 Party Strength

It is more than plausible that organisational assets and other party resources including leadership should be important pre-conditions for RRP success, but in applied research, they are often overlooked, because they are difficult to measure and tend not to vary too much over time. Carter (2005) is one of the very few studies that systematically incorporates party strength into a quantitative model of Radical Right support. Distinguishing between “(1) weakly organised, poorly led and divided parties, (2) weakly organised, poorly led but united parties, and (3) strongly organised, well-led but factionalised parties” she finds that the latter group performs substantially better than the former two (Carter, 2005, pp. 98-99).

David Art’s qualitative study of Radical Right party organisations in twelve West European countries (Art, 2011) provides an important complement to this finding. Taking a longitudinal perspective, Art shows that prospective RRPs need to attract ideologically moderate, high-status activists early in the process to build sustainable party structures and become electorally viable. Otherwise, there is a high probability that they will be subject to factionalism and extremism, which renders them unattractive for most voters.

While Art and Carter compare parties and countries, it is also possible to incorporate information on organisational strength in a within-country model of Radical Right voting. Erlingsson, Loxbo, and Öhrvall (2012) identify a positive effect of “local organisational presence” on the vote of the Sweden Democrats in the 2006 and 2010 elections. One the one hand, this modelling strategy is advantageous, because it maximises the number of cases and can avoid aggregation bias. On the other hand, the validity of Erlingsson, Loxbo, and Öhrvall’s findings is threatened by endogeneity: parties will be more inclined to invest resources and prospective activists will be more inclined to create and join a local organisation if there is a prospect of success in the first place.

3.2 Party Ideology

As a general rule, RRPs take political positions that are in some ways more radical than what the mainstream right is offering, but the ideological heterogeneity of the RRPs is sometimes baffling. It took therefore more than a decade to establish some sort of consensus that these parties do indeed form a party family (Mudde, 1996), and twenty years down the line, scholars still find it difficult to agree on a name for this family, although “Radical Right” is arguably the most popular label at the moment. There are various attempts to distinguish between subgroups within this large cluster. Mudde (2007) identifies a small number of parties that he classifies as “Extreme Right”, i.e. aiming at replacing democracy with some authoritarian system. Similarly, Golder (2003b) draws a line between “populist” and “neo-fascist” parties. Summarising electoral data from Western Europe for the 1970-2000 period, Golder (2003b, p. 444) notes that support for the “neo-fascist” group was very limited in the first place and further declined over time, whereas the appeal of the “populist” parties has grown enormously since they emerged in the 1980s. By and large, this finding still holds today: In Western Europe, where democracy has become “the only game in town”, the vast majority of voters deems openly non-democratic parties unelectable.4 In other European countries where democracy is newer, however, even overtly extremist parties may be electorally successful (see Ellinas 2013; Ellinas 2015 for Greece, Mudde, 2005 and Mareš and Havlík, 2016 for Central and Eastern Europe after 1990, and Stojarová, 2012 for former Yugoslavia).

A different classification, which is not based on the fundamental question of support for democracy but rather on policy positions, was developed by Herbert Kitschelt in his seminal monograph (Kitschelt, 1995). Kitschelt aims at locating RRPs in a policy space that is spanned by two dimensions: a purely economic left-right axis (state vs market) and a more complex dimension that encompasses issues of citizenhood (“group”, see Kitschelt, 2013) on the one hand and individual and collective decision making (“grid”) on the other. Originally, Kitschelt claimed that the then unusual blend of market-liberalism and authoritarian social conservatism represented an “electoral winning formula”. While this may still hold in the US, RRP voters in Western Europe are no longer interested in market liberalism (Lange, 2007; Arzheimer, 2009b), if they ever were. Moreover, electorally successful RRPs have recently de-emphasised their positions on the “grid” (authoritarian) dimension (Kitschelt, 2013, see also section 2.3.2).

3.3 Party System Factors

RRPs do not operate in a vacuum. While they may have a degree of control over their leadership/candidates, their organisational structure, and their ideology, they are but one part of the larger party system, and the words and actions of other parties may have as big an impact on the Radical Right’s electoral fortunes as anything that the RRP themselves do. Presumably, there are two major and partly competing mechanisms at work: From a Downsian logic, it follows that a successful RRP will eventually emerge if there is a demand for more restrictive (migration) policies, which is not satisfied by the existing parties in general and the mainstream right in particular. In this view, a mainstream right party that is soft on immigration and/or the existence of a formal “Grand Coalition” between centre-left and centre-right parties will have a positive impact on the Radical Right vote.

The psychological counter-argument is that political demands are rarely fixed, and that an elite consensus to de-emphasise immigration as a political issue (Zaller, 1992) and to impose a cordon sanitaire might rob the Radical Right of its potential support. Whether this latter strategy is politically feasible is quite a different question. Centre-right parties may have strong incentives to shore up the Radical Right in a bid to strengthen the rightist bloc (Bale, 2003). Centre left parties may want to split the right-wing vote: Mitterand’s decision to hold the 1986 legislative election under PR and Kreisky’s kind words for Haider are cases in point.

The empirical evidence is somewhat mixed. Arzheimer and Carter (2006) find no statistical effect of the mainstream right’s ideological position, or of ideological convergence between the centre left and centre right, but note a substantial positive impact of Grand Coalitions. This result, however, may be shaped by the inclusion of respondents from Austria, which features a long and almost unique history of Grand Coalitions and a consistently strong RRP. On the other hand, Lubbers, Gijsberts, and Scheepers (2002) report that a restrictive “immigration climate” (operationalised as the vote-share weighed average of the other parties positions on immigration) increases the likelihood of a Radical Right vote. Using a slightly different approach that is derived from Zaller’s work, Arzheimer (2009a) notes that the Radical Right benefits from an increasing salience of their issue, regardless of the direction of the statements, and Dahlstroem and Sundell (2012) find a positive effect of anti-immigrant positions held by local politicians from other parties. Again, endogeneity could potentially be a problem in these studies, although this seems less likely in the case of data based on an expert survey (Lubbers, Gijsberts, and Scheepers, 2002) or party manifestos (Arzheimer and Carter, 2006; Arzheimer, 2009a).

3.4 Social Capital

In line with classic theories of the “mass society” (Kornhauser, 1960; Bell, 2002), the rise of the Radical Right has sometimes been linked to widespread feelings of isolation and Anomia. If this relationship holds, higher levels of Social Capital (Putnam, 1993) should curb support for the Radical Right.

Once more, the empirical evidence is limited and contradictory. In a series of case studies in Western and Eastern Europe, Rydgren (2009; 2011) finds that membership in civic organisations does not reduce the probability of casting a vote for the Radical Right. But this does not necessarily disconfirm the Social Capital hypothesis, because Social Capital is not an individual-level but rather a meso-level concept. Coffé, Heyndels, and Vermeir (2007), on the other hand, demonstrate in their model of RRP voting in Flanders that the Vlaams Blok performs significantly worse in municipalities with higher levels of associational life, ceteris paribus, but this finding might be the result of aggregation bias as the authors rely exclusively on census data and electoral counts. Finally, Fitzgerald and Lawrence (2011) combine micro and meso data to estimate a multi-level model of support for the Swiss People’s Party. Even after controlling for a host of variables at the person and at the “commune” level, they find that a municipality’s “social cohesion index” has a substantial positive effect on the probability of a vote for the Radical Right. But while their research design and statistical model are close to ideal, it is not quite clear what they actually measure. Their index includes the proportion of the working population who are not commuters, the proportion of residents who speak the most common language in a given municipality, and the percent of residencies inhabited by their owners. These variables may relate to “bonding” Social Capital, which could explain the positive effect on the RRP vote, but further research is clearly needed.

4 Macro-level Factors

4.1 Institutional Factors

The impact of institutional factors – most prominently, features of the electoral system, decentralisation, and welfare state protection – are very difficult to assess, because they change very slowly or not at all over time and are hence highly correlated with any idiosyncratic unit (=country) effects. Somewhat unsurprisingly, empirical findings are mostly contradictory and inconclusive. As regards electoral systems, Jackman and Volpert (1996) claim that the Radical Right benefits from lower electoral thresholds, but Golder (2003a) argues that this conclusion is based on an erroneous interpretation of an interaction effect and a somewhat idiosyncratic data collection effort. In the same vain, Carter (2002) reports that electoral support for the Radical Right is unrelated to the type of electoral system that is in place in a given election, whereas Arzheimer and Carter (2006) find a positive effect of more disproportional systems but maintain that this might be an artefact.

As regards features of the welfare state, Swank and Betz (2003) find that higher level of welfare state protection seem to reduce the appeal of the Radical Right. However, their analysis is based exclusively on macro data. Using a more specific indicator (generosity of unemployment benefits) and micro data, Arzheimer (2009a) finds that more generous benefits, which may cause “welfare chauvinism”, are linked to higher levels of support but only if levels of immigration are below average (see also next section).

4.2 Immigration and Unemployment

For obvious reasons, the two macro-level variables whose effects have been most extensively studied are immigration, unemployment, and their interaction: a high immigration / high unemployment situation represents perhaps the most clear-cut scenario for ethnic competition for scarce jobs. Nonetheless, the findings are far from conclusive, as can be seen by looking at two of the first comprehensive comparative studies: While Jackman and Volpert (1996) find a substantial positiveeffect of aggregate unemployment on the Radical Right vote, Knigge (1998), who uses a design that is quite similar, reports a negative effect. So do Arzheimer and Carter (2006). Lubbers, Gijsberts, and Scheepers (2002), in their first multi-level model of Radical Right voting in Western Europe, find no significant relationship between the unemployment rate the Radical Right voting intentions, whereas Golder (2003b), whose analysis is once more based on aggregate data, reports a positive (main) effect as well as a positive interaction between unemployment and immigration. Finally, Arzheimer’s (2009a) results from a rather complex multi-level model of Radical Right voting suggest that unemployment may have a positive effect under some scenarios when unemployment benefits are minimal and contributing factors (both individual and contextual) are already favourable.

Although measures for immigration are hardly ideal and differ across studies, results for the effect of immigration are less equivocal: Knigge (1998), Lubbers, Gijsberts, and Scheepers (2002) , Golder (2003b), Swank and Betz (2003), and Arzheimer and Carter (2006) all find a positive effect of (national) immigration figures on the likelihood of a Radical Right vote. Arzheimer (2009a) by and large confirms this, although with an important qualification: In his study, the interaction between unemployment and immigration is negative so that a high levels of both variables, their effects do not reinforce each other any more but rather hit a ceiling. Moreover, generous unemployment benefits reduce the effect of immigration.

4.3 Crime

Like immigration and unemployment, high crime rates are supposed to benefit the Radical Right, but there is not much empirical evidence to back up this claim. Coffé, Heyndels, and Vermeir (2007) conducted one of the first studies that tests the alleged relationship. In an aggregate model of Vlaams Blok support in Flemish municipalities, they find that high crime rates increase the likelihood of the Vlaams Blok contesting an election, presumably because the party anticipates higher levels of support. However, once this selection mechanism is accounted for, crime has no positive effect on the Vlaams Blok’s result.

The study by Coffé, Heyndels, and Vermeir has three distinct advantages: It models the decision to compete in an election and the results of that decision separately, it is built on a large number of cases, and the level of aggregation is low. But unfortunately, their design does not allow for comparisons across time or political systems. In a sense, the article by Smith (2010) provides the complement to their work: Smith studies the relationship between support for the Radical Right and crime rates at the highest possible level of aggregation by analysing 182 national parliamentary elections that were held in 19 Western European countries between 1970 and 2005. Controlling for unemployment, inflation, immigration, and various interactions, he finds that higher crime rates are associated with stronger support for the Radical Right. This relationship becomes stronger if immigration rates are higher.

Finally, the contribution by Dinas and Spanje (2011) specify a multi-level model of Radical Right voting in the Netherlands in 2002. Like in the case of Coffé, Heyndels, and Vermeir (2007), their results are confined to one election in a single country. As they combine individual and contextual data, there is no aggregation bias, and they can even tease apart the effects of objective crime rates and subjective attitudes towards crime. Their results suggest that the effects of crime and immigration do not operate across the board but rather only affect those citizens who perceive a link between the two.

4.4 Media

One final variable at the macro level that attracts considerable interest is the media coverage of the Radical Right’s issues. While voters will be exposed to crime, immigration and unemployment to one degree or another, media reports may have a stronger effect than personal experiences or non-experiences via two alleged mechanisms: Theories of agenda setting claim that the media, by focusing on certain topics, select a handful of politically relevant issues from a much larger pool of problems. Those issues on the agenda then serve as yardsticks for evaluating parties, an effect known as priming (Scheufele and Tewksbury, 2007). In extreme cases, an issue may become so closely associated with a party that this party “owns” the issue (Petrocik, 1996) and will almost automatically benefit whenever it achieves a high rank on the agenda. Green parties and the environment are an oft-cited example, but the Radical Right and immigration have become a close second in the eyes of many observers (Meguid, 2005).

Notwithstanding the importance of the alleged nexus between media coverage and Radical Right support, the evidence is limited once more. The main reason for this is that data on media content are difficult to come by and expensive to produce in the first place. This is slowly changing now, with automated coding methods and open data bases such as GDELT providing new avenues for research, but even so, matching media with micro-level data is next to impossible, because mass opinion surveys do not normally collect detailed (i.e. per item) information on media consumption. Most of the existing research is therefore based on aggregated (i.e. time-series) data.

In their pioneering study, Boomgaarden and Vliegenthart (2007) find a positive relationship between salience of immigration in Dutch media and aggregate support for Radical Right parties during the 1990-2002 period, net of any changes that can be ascribed to the unemployment and immigration rates and their interaction. This article is complemented by Koopmans and Muis (2009), who focus on the end of that period (i.e. Pim Fortuyn’s 2002 campaign) and aim to identify a number of “discursive opportunities” that facilitated Fortuyn’s breakthrough. In another study that resembles their 2007 piece (Boomgaarden and Vliegenthart, 2009), Boomgaarden and Vliegenthart can further demonstrate a link between news content and anti-immigrant sentiment in Germany for the 1993-2005 period.

Finally, in a bid to overcome the dearth of micro-level data on media consumption from mass surveys as well as the limits of the ex-post-facto design, interest in in experimental studies has grown considerably over the last decade. One such study is that by Sheets, Bos, and Boomgaarden (2015), who exposed members of an online-access panel to an synthetic news article. Some small parts of this article were systematically varied to provide “cues” that would prime the issues of immigration, anti-politics, and the RRP itself. While Sheets, Bos, and Boomgaarden can demonstrate some effects of these cues on anti-immigrant attitudes, political cynicism, and ultimately on PVV support, some question marks remain. First, the effects on anti-immigrant attitudes are weak compared to those on political cynicism. Second, like with any experimental intervention, it is not clear if effects of a similar magnitude occur “in the wild”, and if so, how long they persist. Third, the experiment was designed in a way that means that the immigration and anti-politics cues were always combined with an RRP cue, which will in all likelihood bias the estimates for their respective effects either upwards or downwards. Clearly, further (cross-national) research is needed.

5 Small Area Studies

By now it should be clear that nearly all authors in the field treat support for the Radical Right as a multi-faceted phenomenon that must be explained at multiple levels, with unemployment, immigration, and political factors and media cues being the most prominent contextual variables. Most studies measure these variables at the national level, but living conditions in European states vary considerably across regions, so designs that compare provinces, districts or even neighbourhoods within countries are becoming more and more prominent. One of the first of these studies was conducted by Bowyer (2008), who looks at electoral returns for the British National Party (BNP) in several thousand wards in the 2002/2003 local elections in England. He finds that the BNP was strongest in predominantly white neighbourhoods that are embedded within districts which are characterised by the presence of large ethnic minorities, a pattern that has been described as the “halo effect” (Perrineau, 1985). Economic deprivation (though not necessarily unemployment) also played a role. Similarly, Rydgren and Ruth (2011), who analyse support for the Sweden Democrats in the 2010 election across the country’s 5668 voting districts, show that the party did better in poorer districts with bigger social problems. Once these factors are controlled for, there is also some evidence for the existence of a “halo effect”.

Other studies have focused on units that are larger but politically more meaningful than census districts or electoral wards, e.g. departements, provinces, or sub-national states (Kestilä and Söderlund, 2007; Jesuit, Paradowski, and Mahler, 2009), accepting possible aggregation bias in exchange for the ability to include political and/or media variables in the model. The former study reports positive effects of unemployment and some institutional variables but no effect of immigration, whereas the latter identifies some complex interactions that link immigration and unemployment to Radical Right support via an increase in inequality and a lack of social capital.

Studies in small(ish) areas are currently one of the most promising avenues of research into the Radical Right vote, be it on the level of subnational political units or in even smaller tracts. Either way, researchers need to account for the fact that an increasing number of voters are either immigrants or the offspring of immigrants, who will be disinclined to support the Radical Right. Estimates from small area studies that are based on aggregate data will therefore be biased downward (Arzheimer and Carter, 2009b). Hence, multi-level analyses that combine micro data with information on local living conditions are the way forward in this particular branch of research.

6 Conclusions

Over the last three decades, Radical Right parties have become a permanent feature of most European polities. Their rise, persistence, and decline can be quite well explained by the usual apparatus of electoral studies. On the micro level, the most important factors are value orientations, attitudes towards social groups, candidates and political issues as well as (the lack of) party identifications. At the macro level, social change (broadly defined) undoubtedly plays an important role, while parties, the media and all other sorts collective actors operate at the meso-level in between.

Because RRPs are often perceived as divisive, disruptive, or outright dangerous, a great deal of intellectual energy has been spent looking for “deeper” explanations. And indeed, there can be very little doubt that the presence or absence of immigrants and immigration, the frequency and nature of contacts between the immigrants and the native population, and the way immigration is framed by other political actors and the media is a major contributing factor to Radical Right support. However, given that immigration, ethnic tensions, and RRP actors are almost ubiquitous in Western societies, their success is not a major surprise. Ultimately, trying to understand why they are not successful in some cases might be more rewarding, both politically and intellectually.

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Sheets, Penelope, Linda Bos, and Hajo G. Boomgaarden (2015). “Media Cues and Citizen Support for Right-Wing Populist Parties”. In: International Journal of Public Opinion Research. DOI: 10.1093/ijpor/ edv014.

Smith, Jason Matthew (2010). “Does Crime Pay? Issue Ownership, Political Opportunity, and the Populist Right in Western Europe”. In: Comparative Political Studies 43.11, pp. 1471–1498. DOI: 10.1177/ 0010414010372593.

Stojarová, Věra (2012). “The extreme right in Croatia, Bosnia-Herzegovina and Serbia”. In: Mapping the Extreme Right in Contemporary Europe. From Local to Transnational. Ed. by Andrea Mammone, Emmanuel Godin, and Brian Jenkins. London et al.: Routledge, pp. 143–158.

Sumner, William Graham (1906). Folkways. A Study of the Sociological Importance of Usages, Manners, Customs, Mores, and Morals. Boston: Ginn.

Swank, Duane and Hans-Georg Betz (2003). “Globalization, the Welfare State and Right-Wing Populism in Western Europe”. In: Socio-Economic Review 1, pp. 215–245.

Swyngedouw, Marc (2001). “The Subjective Cognitive and Affective Map of Extreme Right Voters: Using Open-ended Questions in Exit Polls”. In: Electoral Studies 20, pp. 217–241.

Taggart, Paul (1995). “New Populist Parties in Europe”. In: West European Politics 18.1, pp. 34–51.

Tajfel, Henri and John C. Turner (1986). “The Social Identity Theory of Intergroup Behaviour”. In: Psychology of Intergroup Relations. Ed. by Stephen Worchel and William G. Austin. Chicago: Nelson-Hall Publishers, pp. 7–24.

Taylor, C. Marylee (2002). “Fraternal Deprivation, Collective Threat, and Racial Ressentment”. In: Relative Deprivation. Specification, Development, and Integration. Ed. by Iain Walker and Heather J. Smith. Cambridge: Cambridge University Press, pp. 13–43.

Vasilopoulou, Sofia (2011). “European Ingegration and the Radical Right. Three Patterns of Opposition”. In: Government and Opposition 46.2, pp. 223–244.

Walker, Iain (2001). “The Changing Nature of Racism: From Old to New?” In: Understanding Prejudice, Racism, and Social Conflict. Ed. by Martha Augoustinos and Katherine J. Reynolds. London, Thousand Oaks: Sage, pp. 24–42.

Werts, Han, Peer Scheepers, and Marcel Lubbers (2013). “Euro-scepticism and radical right-wing voting in Europe, 2002 2008: Social cleavages, socio-political attitudes and contextual characteristics determining voting for the radical right”. In:European Union Politics 14.2, pp. 183–205. DOI: 10.1177/1465116512469287.

Zaller, John R. (1992). The Nature and Origin of Mass Opinion. Cambridge, New York, Oakleigh: Cambridge University Press.

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1At least at the attitudinal level, old and modern racism seem to be closely related (Walker, 2001).

2Although value orientations are sometimes grouped together with personality traits, they will be discussed in a separate section below.

3Similar arguments have been made about the rise of the right-wing extremist movements in the 1920s as well as about their resurgence in the postwar years (e.g. Scheuch and Klingemann, 1967).

4Marine Le Pen’s attempts to soften the image of the Front National (Mayer, 2013) and her public clashes with her father over his unreformed anti-semitism are a case in point.

The AfD: Finally a Successful Right-Wing Populist Eurosceptic Party for Germany?

 

Germany is unusual amongst West European countries because all relevant parties (with the possible exception of the Left party) are unwavering supporters of European integration. Moreover, while the Radical Right is now a permanent feature a of many European democracies, the electoral successes of Germany’s Radical Right parties have been very modest and confined to the subnational level.

However, in 2013, only months before the General Election, a new party was formed that campaigned for a dissolution of the Eurozone and a radical re-configuration of German foreign policy. This new “Alternative for Germany” (Alternative für Deutschland or AfD for short) came tantalisingly close to the electoral threshold of five per cent. Nine months on, the party polled seven per cent in the 2014 European parliamentary election and was eventually admitted to the European Conservatives and Reformists group (ECR), which further soured the relationship between German Chancellor Angela Merkel and British Prime Minister David Cameron. In three (eastern) state parliamentary elections held in August/September 2014, the AfD did even better, capturing between 9.7 (Saxony) and 12.2 (Brandenburg) per cent of the vote.

The AfD has been described as eurosceptic and right-wing populist1 by its political rivals and by the mainstream media. If this description was correct, it would signal a qualitative shift in the structure of party competition in Germany. Moreover, due to such a party’s blackmail potential vis-a-vis the moderate right, this would constitute a massive shock to the German party system, with considerable implications for Germany’s future position on European integration and for German immigration policies. It is, however not at all clear if and to what degree such a classification of the AfD is warranted, as those terms are used rather indiscriminately in mediated discourses (Bale, Kessel, and Taggart 2011).

The aim of this article is therefore simply to assess the AfD based on categories derived from the rich comparative literature on the Radical Right and on euroscepticism. Since the AfD is a very young party with no parliamentary record, the primary source of evidence the party’s 2014 European manifesto. Additional information is drawn from material on the party’s website and Facebook presence.

The remainder of this article is organised as follows: The next section briefly reviews the concepts that will be used in the analyses. The third section provides some background information on euroscepticism and right-wing radicalism in Germany, and on the short career of the AfD. The fourth section presents an in-depth analysis of the AfD’s manifesto and other texts produced by the party. The final section summarises the main findings and puts them into perspective.

Concepts

Radical right-wing populism

In the early 1980s, a new group of right-wing parties emerged in Western Europe. These parties differed significantly and systematically from mainstream parties of the right and were therefore portrayed as a new party family in the scholarly literature. While there is little disagreement as to which parties belong to this new family, research on these parties and their voters has been plagued by the twin questions of what exactly sets these parties apart from the mainstream right, and what adjectives (“radical”, “populist”, “extreme”, “anti-immigrant” …) best capture these differences.

More recently, Mudde (2007) has proposed a new scheme for classifying right-wing parties outside the mainstream that has won international acclaim because it accommodates a wide range of parties while identifying important differences between them. According to Mudde (2007, 19), the lowest common denominator for the party family is “nativism”, an ideology that combines nationalism and xenophobia. Nativism is a broad concept that subsumes racism, ethnocentrism, and anti-immigrant sentiment. Nativism holds that non-native elements (persons, ideas, or policies) present a threat to the nation state, which should be as homogeneous as possible.

However, traces of nativism may be found within the manifestos of mainstream parties. Following Mudde (2007, 21–23), to qualify as a Radical Right, a party additionally needs to display authoritarian tendencies, i.e. an aggressive stance towards political enemies and a preference for a strictly ordered society, strong leadership, and severe punishments for offenders. With authoritarianism comes a political bent that is not necessarily anti-democratic per se but goes against the grain of some of the fundamental values and principles of liberal democracy (Mudde 2007, 25–26) such as tolerance, pluralism, and the protection of minorities and their rights.

Within the Radical Right, Mudde then identifies a subgroup of parties that is also populist in nature. By populism, Mudde (2007, 23) means not just a style of political communication but rather a “thin ideology” (see Stanley 2008) that pits the “pure people” against a corrupt elite and puts majority rule above human rights and constitutional checks and balances. This Populist Radical Right is arguably the most electorally successful subtype within the larger party family.

Finally, a small (and not necessarily populist) subgroup of the Radical Right is actually anti-democratic. Borrowing from the long-standing practice in Germany, Mudde labels these parties as “Extreme Right”.

While Mudde’s system of definitions may not have ended the debate about terminology in the field, it obviously provides a useful tool for assessing new parties, and more generally for discussing developments in Germany within a wider European context.

Euroscepticism and ideology

Euroscepticism broadly refers to a negative stance towards European integration. As a field of scientific inquiry, it only took off in the late 1990s (see Vasilopoulou 2013) when the “Post-Maastricht Blues” (Eichenberg and Dalton 2007) kicked in. Mudde (2012) distinguishes between two main strands in this literature: the “North Carolina school”, which clusters around the Chapel Hill dataset, and the “Sussex school”, which chiefly relies on case studies of party manifestos. Whereas the “North Carolina school” aims at quantifying degrees of euroscepticism, the Sussex group introduced a qualitative distinction between “hard” and “soft” euroscepticism. “Hard” euroscepticism refers to a principled rejection of European integration that is ultimately incompatible with EU membership. “Soft” euroscepticism is not opposed to integration as such, but rejects the current state of European politics as well as the trajectory towards an “ever closer union” (Szcerbiak and Taggart 2008, 1:7–8).

Mudde (2012, 194) notes that in the face of low salience of Europe and euroscepticism, it can be surprisingly difficult to determine whether a party is soft eurosceptic, hard eurosceptic, or not eurosceptic at all. However, in the case of the AfD, which emerged as an anti-Euro party and drew up their first full-length manifesto in the context of the 2014 European election, there is clearly no lack of salience. The hard/soft distinction is therefore a useful template for the analysis of the AfD’s ideology.

But it is not entirely obvious how euroscepticism relates to broader ideologies. In most West European polities, parties position themselves within a two-dimensional (Kitschelt 1995; Benoit and Laver 2006) or perhaps even three-dimensional (Bakker, Jolly, and Polk 2012) space. The everyday language of politics, however, still relies on the traditional left-right-dichotomy, and most West Europeans are quite happy to place themselves on a unidimensional left-right-scale (Lo, Proksch, and Gschwend 2014). While the precise meaning of “left” and “right” may vary across time and space (Huber and Inglehart 1995), left-right-placement usually reflects the perceived distance between voters and parties as well as value-based preferences (Knutsen 1997). The focal value on the right-hand side of this spectrum is inequality (that can be the result of some “natural order”, mandated by tradition and authority, or the product of some market mechanism). The most important value on the left-hand side is equality (see e.g. Bobbio (1997)), realised through state regulation and redistribution of resource that may or may not infringe on liberty and property rights.

Data reported in Ray (1999)’s (1999) early seminal contribution suggest that eurosceptic parties were by and large located at both ends of the political spectrum during the 1990s. The analysis by Marks et al. (2006), which draws on more recent data, confirms precisely such a curvilinear pattern, at least for Western Europe. But even at the very (right) extreme of the political spectrum, euroscepticism is neither omnipresent nor universally “hard” (Vasilopoulou 2013). Therefore, a new party such as the AfD must be very carefully evaluated before it can be classified as right-wing, populist, eurosceptic, or all of the above.

Conditions for radical right success: Demand, context, and supply

Support for Radical Right parties varies considerably across time and political systems. The burgeoning literature has identified three groups of factors that can help to make sense of this variation: “Demand-side” variables refer to individual features such us gender, formal education, class, and, most importantly, political disaffection and anti-immigration attitudes (Brug and Fennema 2007) that make voters more or less susceptible to right-wing mobilisation. Their effect is moderated by contextual conditions, which include institutional factors (e.g. the electoral system or the degree of political centralisation), socio-economic conditions (e.g. the unemployment rate or the annual number of new asylum applications), and political variables such as the salience of the immigration issue for other parties and the media, or the willingness of the elites to co-operate with the Radical Right (Arzheimer 2009).

Demand-side and contextual variables collectively form the external environment to which a Radical Right party has to adapt, at least in the short term.2 A third group of variables, however, is more or less under the control of the party. Such “supply-side” factors include the party’s policy proposals, candidates for office, and general public appearance.3 Amongst these, past research has highlighted the availability of a “charismatic leader” as a precondition for Radical Right success, although this hypothesis is highly contested (Brug and Mughan 2007).

More recently, David Art (2011) has developed a more nuanced account that stresses the importance of party activists in general. In a nutshell, Art argues that the trajectory of Radical Right parties hinges on the nascent party’s ability to attract a sufficient number of the right type of activists, which in turn depends on historical legacies and the initial reaction of mainstream political actors to the new party (Art 2011, 31). To have a chance of electoral success, a new party needs many “moderates”, i.e. nationalists who credibly subscribe to the rules of liberal democracy and steer clear of biological racism and neo-nazism. Ideally, these moderates should also have high social-economic status (SES) and a degree of political experience outside the Radical Right (Art 2011, 33). Conversely, the emerging party should try to curb the number of “opportunist” members without strong political convictions and to avoid attracting any “extremist” activists who are openly hostile to parliamentary democracy: The latter group is prone to infighting over (highly fragmented) political principles, unwilling to temper their political demands in order to appeal to more moderate voters, and provide an easy target for any attempts to ostracise the new party. Moreover, they are often unexperienced (Art 2011, 35–40).

Whilst this article is chiefly concerned with the question of whether the AfD can at all be classified as Radical Right, Art’s theory of radical right mobilisation provides a useful template for the next section and puts the AfD’s electoral appeal in perspective.

Euroscepticism and the radical right in Germany

The lack of successful eurosceptic and right-wing parties

German MPs generally support European integration and even subscribe to a “deep core belief” of “the EU as a good thing” (Kropp 2010, 140). Almost all parliamentary parties in Germany are staunch proponents of European political integration. On the right, both the FDP and the CDU/CSU have supported and shaped European integration from its inception in the 1950s, although the small Bavarian CSU has been occasionally been more critical of the commission and some European policies than its sister party. On the left, the SPD and the Greens have taken a similar stance, both in opposition and in government (Wimmel and Edwards 2011, 295–296).

Only the Left party have voted consistently against the treaties of Maastricht, Nice, and Lisbon, because they reject the “neo-liberal” Single European Market, the monetary and budgetary policies mandated by the Stability and Growth Pact, and the “militaristic” Common Foreign and Security Policy. But even the Left party have declared themselves pro-European in principle at these occasions (Wimmel and Edwards 2011, 306–308), which makes them soft eurosceptics.

Otherwise, German euroscepticism has been confined to a number of unsuccessful single-issue fringe groups such as the Pro-Deutschmark party, and to the country’s three Radical Right parties (Lees 2008): the Republicans, the DVU, and the NPD. While all three had occasional successes in state elections and the Republicans even were represented in the 1989-1994 EP, their support proved fickle, and neither of them has ever won representation in the Bundestag. Compared to other West European countries, this weakness of the Radical Right appears anomalous and makes Germany a large negative outlier in a statistical model of radical right voting in Western Europe that controls for demand-side and contextual factors (Arzheimer 2009).

Art’s assessment of Radical Right mobilisation in Germany (Art 2011, 190–208), however, provides a plausible explanation for this lack of right-wing success. German elites have stigmatised National Socialism and criminalised the use its symbols very early on whilst offering nationalist a home in the mainstream centre-right. This strongly discouraged ‘moderates’ and ‘opportunists’ from joining the NPD, DVU, or the Republicans. These parties in turn have always had a fixation with the past (Ignazi 1992; Kitschelt 1995), which ruled them out as serious political players and made it easy to create and maintain a cordon sanitaire between them and the main parties.

Against this backdrop, the meteoric rise of the AfD and its ability to steer clear of any Nazi connotations is a very unusual4 and significant development. Arguably, this success was only possible because the party was formed by “moderates” with very high SES, considerable civic skills, and some political experience (detailed in the next seciont), and rests on the party’s continuing ability to ward off “extremists”, or at least activists that are perceived as too extreme in the German context. This makes the question of the AfD’s classification all the more pertinent.

The making of the “Alternative for Germany”

The AfD began its political life in September 2012 when a group of disaffected CDU members including Konrad Adam (born 1942), Alexander Gauland (born 1941), and Bernd Lucke (born 1962) founded a political action group called “Wahlalternative 2013” (an electoral alternative for the 2013 General election). While none of them played a leading role in the CDU, all three had been party members for several decades and were reasonably prominent figures: Adam and Gauland are well-known conservative journalists, while Lucke is a professor of economics who has been instrumental in organising two petitions by academic economists against the various bailout packages.

However, the AfD should not be considered a splinter party from the CDU, because the founding members were recruited from a broader centre-right background: Other signatories included 28 university professors (almost all of them economists), entrepreneurs and managers, and a former state party chair of the FDP (the Liberal party). The Wahlalternative’s short manifesto5 demanded that Germany should not guarantee any foreign sovereign debt, that all members of the Eurozone should be free to re-introduce national currencies or to join new currency unions, and that any further transfer of German sovereignty should be subject to a referendum.

Initially, the Wahlalternative was organised as a pressure group that supported the “Federation of Independent Voters”, a fledgling umbrella organisation for community-based, voter associations that are often dominated by the owners of small local business. In January 2013, both organisations jointly drew up a slate of candidates for the state election in Lower Saxony. However, the list polled just over one per cent of the vote, much less than the five per cent required for parliamentary representation. Subsequently, the two groups parted ways, and in February 2013 the Wahlalternative’s leadership formally founded the AfD as a political party, with the stated intention to run in the upcoming federal election on September 22. Adam, Lucke, and Frauke Petry (born 1975), a chemist and entrepreneur from the eastern state of Saxony, were elected to jointly lead the party.

By July, the party had drawn up a short manifesto that focused on monetary and fiscal policies, had set up branches in all 16 Länder, and had attracted more than 10,000 members.6 In the end, the AfD garnered 4.7 per cent of the vote, the best result for any party competing for the first time since 1953. While the AfD narrowly missed the electoral threshold, this result was widely seen as a remarkable achievement that gained them a foothold in the political system and gave them access to state funding.

Over the following six months, the party focused on broadening their programmatic profile and shedding the image of the single-issue party. During this time, it became clear that there was considerable potential for conflict within the party. In some state-level branches, the leadership resigned or was ousted over allegations of financial, political, or personal misconduct.7 More importantly, it became clear that various factions (conservatives, liberals, right-leaning Christian Democrats and perhaps even Christian fundamentalists) were warring for influence within the party. In January/February 2014, a party conference that was supposed to select the candidates for the European election had to be suspended for a week, because the delegates could not agree on a slate. In March, another party conference rejected a change to the statutes that would have made it possible for Lucke to become sole party leader.8 Lucke barely managed to take control of a debate on the party’s position on homosexuality (started by himself) and struggled to enforce a party line that stops short of open populism and hard euroscepticisim.

The most visible split within the party concerned the question of its future membership in a political group in the European Parliament. While Lucke was adamant that the AfD should join the ECR, some of the party’s rank-and-file and the party’s youth organisation “Young Alternative” would rather have worked with the Europe of Freedom and Democracy (EFD) group. Things came to a head when the Young Alternative invited Nigel Farage to give a lecture in Cologne.9 Lucke intervened but could neither forestall the event, nor was he successful in reprimanding Marcus Pretzell, one of the organisers and also a member of the party’s executive committee and a candidate for the EP election.

Electorally, none of this did the party any harm. In the polls, support for the AfD had been consistently in the range of six to eight per cent. In the actual election, they won 7.1 per cent of the vote, which entitled them to seven seats in the European Parliament – as many as the Left party and more than the CSU or the FDP have won. The list of elected candidates includes only two women and reflects the bourgeois background of the party leadership.10 On June 12, the seven joined the ECR, making it the third-largest faction in the EP.

An analysis of the “Alternative”

A quantitative analysis of the AfD’s 2013 European election manifesto

Even for European elections, German parties tend to formulate detailed manifestos that cover a lot of policy domains. The AfD is no exception to that rule. The 2014 European election manifesto is the party’s first full-length policy document and therefore very well suited for assessing the party’s official policy positions. To provide context for its analysis, the manifestos of the main parties, the leftist Pirates, and the right-wing extremist NPD were analysed, too.

At 4,894 words, the AfD’s manifesto is close to the median length of 5,852 words. A number of function words (articles, conjunctions, prepositions etc.) were removed from the files. Running headers or footers, tables of contents and adverts were also discarded, but preambles and prefaces by the party leaders were retained. Because German is an inflected language, the “Snowball” stemming algorithm was applied to prepare the texts for quantitative analysis. Stemming aims at reducing words to their roots by removing suffixes and affixes so that different inflected forms are grouped together as a single item. While stemming is less accurate than full lemmatisation (determining the dictionary form of inflected words), it can be carried out quickly and efficiently to reduce the complexity of a text, and the loss in precision does not matter much in practical applications (Grimmer and Stewart 2013, 272).

Although a number of very common German words had been discarded in the first step, some stems such as “Europ” and “EU” appear very frequently in all manifestos and are thus not useful for discriminating between parties. Therefore, the one per cent most frequent stems were removed. Following Grimmer and Stewart (2013, 273), very rare stems that collectively make up one per cent of the total corpus as well as stems that were exclusively used by a single party (typically the party name) were also disregarded.

Party

1st

2nd

3rd

4th

5th

Left

regional

work

combine

ecological

society

Green

ecological

human rights

green

Euro

refugee

Pirates

data

oppose

society

access

allow

SPD

allow

work

education

citizens (female)

democratic

CDU

co-operation

worldwide

digital

need

job

FDP

opportunity

freedom

liberal

citizens (female)

responsible

AFD

member state

demand

Euro

eurozone

reject

CSU

Brussels

allow

future

freedom

needs

NPD

(German) people

Brussels

today

foreign

domain

Table 1: The Five Most Frequent words/stems in Nine Election Manifestos

The remaining 4,430 stems give a very clear impression of the AfD’s priorities. Amongst the 15 most frequent concepts in the AfD manifesto are “member states”, “Eurozone”, “ECB”, and “institutions”. None of these words is amongst the top priorities of any other party. However, the analysis also reveals some similarities. “Competition” features prominently in the AfD’s manifesto, but also crops up frequently in the respective platforms of the FDP and the CDU. “Work” is a common concern of the Left party and the SPD, and both Christian Democratic parties frequently talk about “jobs”. Even looking at just the top five words most frequently used by each party gives a good idea of what they stand for (see Table 1).

The observation that the usage of certain words conveys information on ideological proximity and distance between parties has been formalised by Slapin and Proksch (2008), who derive a statistical model that links word frequency to an underlying left-right dimension. Slapin and Proksch (2008) also develop an estimation procedure they call “wordfish”, which recovers ideological positions from political texts and ideological content of words while controlling for differences in the wordiness of political documents and the global distribution of words. Unlike the related “wordscore” method (Laver, Benoit, and Garry 2003), “wordfish” does not require anchor texts and is thus ideally suited for uncovering the positions of new parties relative to a set of more familiar political actors.

For the present analysis, the words and nine parties were simultaneously scaled using version 1.3 of Slapin and Proksch’s wordfish package for the R statistical system. The algorithm converged quickly on the point estimates. 95 per cent confidence intervals were generated by a parametric bootstrap procedure (a method that does not rely on a normal distribution of the estimates) using 500 draws (Slapin and Proksch 2008, 710).11 Again, there were no convergence problems. The words12 most closely tied to left ideology are “Kürzungspolitik” (austerity policies), “erwerbslos” (unemployed), “Altersarmut” (pensioner poverty), “Migrantinnen” (an inclusive and neutral term for migrants), “Sozialcharta” (social charter), “EU-Politik” (EU politics), “Profit” (profit, a more derogatory term than “Gewinn”), “Rüstungsproduktion” (production of arms), “unbefristed” (open-ended, as in open-ended contract), and “nationalistisch” (nationalistic).

The most right-wing words are “fremd” (foreign or strange), “Volk” (the (German) people, in a very emphatic sense), “verhängnisvoll” (fatal or ominous), “einerseits” (on the one hand), “bürgerfern” (removed or insulated from the interests of ordinary citizens), “Asylbewerber” (asylum seekers), “Gender” (as in gender mainstreaming or similar bugbears of the right), “gängeln” (to boss around someone, typically used with reference to the behaviour of bureaucrats), “Bolognaprozess” (the implementation of the Bolgna accord in German Higher Education), and “schleichend” (creeping, typically referring to slow but sinister political change). As this vocabulary reflects both the socio-cultural and the economic dimension of the left-right dichotomy, the scaling displays a high degree of face validity.

Scaling the AfD EP 2014 Manifesto

Scaling of German election manifestos for the 2014 EP election

Estimates for the party positions are very precise (Figure 1). For most parties, the width of the 95 per cent confidence interval is 0.10 points or less on a scale that ranges from -1.69 to 1.26. Crucially, the interval for the AfD is one of the narrowest. The positioning of the parties themselves will be instantly familiar to any student of German politics. The political spectrum is spanned by the Left party on the one hand and the NPD on the other. The Social Democrats (SPD) and the Christian Democrats (CDU) appear in their familiar centre-left and centre-right positions, with the Greens positioned to the left of the SPD and the FDP to the right of the CDU. The Pirates are located between the Greens and the SPD, which again seems plausible.

Both the CSU and the AfD appear to the right of the FDP, slightly closer to the NPD than to the CDU. The confidence intervals for their positions overlap, which implies that they are statistically indistinguishable.13 Lucke has repeatedly claimed that his party is neither left nor right14 and even stated that the AfD represents a new breed of party (“Partei neuen Typs”) at their founding conference15 – a very awkward pun on the Stalinisation of East Germany’s Socialist Unity party in the 1940s. But their manifesto places them firmly at the far right of the political spectrum.

The position of the CSU is perhaps more surprising, because the Bavarian Christian Democrats have been a fixture of German politics since 1945. But the party has nonetheless been described as anti-immigration and (borderline) right-wing populist in the literature (Lubbers, Gijsberts, and Scheepers 2002; Falkenhagen 2013). Former leader Franz-Josef Strauß famously declared that “there must be no democratic party right of the CSU” (Raschke and Tils 2013, 253, my emphasis). More recently, the party has also steered an ambiguous course towards the EU.16 While the content of their 2014 manifesto may already reflect concerns about the emerging competition from the AfD, the document is nonetheless in line with the CSU’s traditional position at the very margin of the established party system.

Is the 2013 manifesto radical, populist, and eurosceptic?

Against this backdrop, the estimates for both the AfD and the CSU are highly plausible. The general left-right measure paints, however, a very broad-brushed picture of the AfD’s political program. Assessing the question whether the AfD is not only on the right but also radical/extremist, populist, and eurosceptic requires a close reading of its manifesto.

In the theory section, “nativism”, i.e. a mixture of nationalism and xenophobia was proposed as a criterion for separating the Radical Right from other right-wing parties. In line with their overall position on the right of the political spectrum, the AfD is certainly unusually prone (by German standards) to display national symbols and to emphasise Germany’s national interest. “Mut zu Deutschland” (roughly translated: dare to stand by Germany) was the title of their manifesto and their main slogan for the EP 2014 campaign. The phrase is still used prominently on the party’s main website, their social media profiles, and in other party material. The slogan alludes to the common right-wing argument that national pride is systematically discouraged in Germany but was deployed in a more specific sense during the campaign: the AfD wants Germany to act more assertively within the European Union.17

The corresponding section, however, is one of the shortest in the manifesto and makes rather modest demands. The AfD blames the member state governments for breaking the treaties (particularly the Stability and Growth Pact), it demands that the EP should launch a public inquiry into the details of bailout measures, and it suggests (without going into details) that Germany should have a greater say within the European institutions. However, the main opponent for the AfD is an unholy alliance between the EU institutions and Germany’s “Altparteien” (old, i.e. established parties – a term the AfD has borrowed from the Green party of the 1980s). But one would be hard-pressed to find any statement that is nationalistic in the usual sense of the term in this or in fact in any other part of the manifesto.

The section on immigration and asylum also strikes a rather conciliatory tone. The AfD subscribes to the principles of free movement and free choice of residence for all EU citizens, although they want to limit benefits (of which they are critical in general) to long-term residents and their offspring. Moreover, the AfD acknowledges the problems brought about by demographic change and supports a point-based immigration regime for non-EU citizens. Finally, the AfD commits itself to a “humane” asylum system, which implies more financial and logistic support for the member states in the South, common standards for accommodation, and labour market access for asylum seekers. Taken together, these positions are not overly restrictive by German standards and do not display any nativist tendencies.

The AfD rejects Turkish EU membership flat-out and mentions “geographical, cultural and historical borders” in this context. But apart from this, and from a single reference to Europe’s “Christian-occidental values”, religion and culture, which are often used as politically acceptable codewords for non-European ethnic groups (Zúquete 2008), are not at all mentioned in the text. Judging by its manifesto, the AfD is therefore not a Radical Right, let alone an Extreme Right party.

Is the AfD populist? If one defines populism as a “thin ideology”, then there is very little in the manifesto that would support such a claim. The AfD is highly critical of “Brussels”, and of the mainstream parties in Germany. They also argue that the ongoing financial crisis was to a large degree caused by irresponsible behaviour of the banks, which should be regulated more tightly. Moreover, they want to improve the democratic legitimacy of the EU in general and demand that future enlargements as well as important decisions on the Euro should be put to a referendum.

But that alone does hardly make them populists. Their manifesto does not contain a single reference to “elites”, the “political class”, or the “eurocrats”. Corruption is mentioned only once, in the innocuous context of the UN’s anti-corruption charter. But even if one opts for a broader, softer definition that primarily treats populism as a style of political communication “that refers to the people” (Jagers and Walgrave 2007, 322) there is nothing in the manifesto that would appear as particularly populist in that sense.

The AfD’s manifesto does not even conform with every day notions of populism that imply appeal to emotions, oversimplification, and a degree of opportunism (Mudde 2004, 542–543). On the contrary: The AfD’s manifesto contains lengthy references to economic theory, is largely written in a rather technical and stilted language and even contains a couple of footnotes that cross-reference political demands to articles in the Treaty on the Functioning of the European Union.

That leaves the issue of euroscepticism. The AfD is clearly not a “hard” eurosceptic party. They are opposed to the currency union in its present form, to current and future bailouts, and more generally to a federal European state. But at the same time, they are committed to the European Union as such and have dropped their erstwhile demand for a return to the Deutschmark from their manifesto. While they want to strengthen the principle of subsidiarity (which was established in the Treaty of Maastricht at the behest of the German Länder), they don’t intend to reduce the EU to a trade bloc. Although they are highly suspicious of secretive intergovernmental co-operation in Justice and Home Affairs, they support the pursuit of a Common Foreign and Security Policy based on lowest common denominator solutions. Taken together, “soft euroscepticism” best describes the political positions articulated in the manifesto.

The Alternative’s internet presence

In Germany, parties are legally obliged to draw up comprehensive manifestos and lodge them with the Federal Returning Officer. These platforms are routinely scrutinised by researchers and the media. The lack of any obviously radical and populist content in the AfD’s manifesto could therefore be misleading. Indeed, a speech delivered by Konrad Adam on June 27, 201318 gives a rather different impression. Adam encourages party members to become “dangerous citziens” (“gefährliche Bürger”) who dare to take on the elites. Politicians of other parties are portrayed as greedy, lazy, and incompetent predators who are after the money of ordinary taxpayers and sell out the national interest to the EU, and the mainstream media help them to cover up.

Other speeches documented on the website, however (four by Lucke, one by Starbatty and one by Henkel) strike a similar, yet clearly more moderate tone. Starbatty and Henkel in particular discuss intricacies of social and economic policy in great detail, while Lucke often focuses on his vision for the further development of the party. None of these speeches could be considered populist or radical.19

To get a more rounded impression of the party’s appeal, it therefore makes sense to analyse the party’s presence on the internet. The party’s main website (http://www.alternativefuer.de) is built with the wordpress platform. It is professionally designed and maintained and currently (July 2014) consists of more than 1,300 unique HTML pages20 plus 62 PDF documents. Many pages contain redundant content because they serve as archives that bring together posts related to a specific author, tag, category, or date of publication. Other pages are simply of an administrative nature (e.g. contact information). The following analysis is therefore restricted to 371 blog-post like pages, which contain comments on media reports, current events, or simply document statements by prominent party leaders.

For the analysis, only text in the main body of the pages was extracted. Stopword removal and stemming were conducted as outlined above. The resulting corpus consists of 45,990 words, which can be reduced to 9,745 stems. Obviously, posts on a website serve a function that is different from that of the manifesto, and this is reflected in the language used. A simple count demonstrates that the AfD is chiefly talking about itself and its leadership. “AfD” (573), “Alternative” (338), and “Deutschland” (“Germany”, 531) are amongst the five most popular stems. Also very prominent is the name of Bernd Lucke (274), who clearly overshadows his fellow leaders Adam (42) and Petry (48). Far more important than Adam and Petry are deputy leaders Gauland (151) and Henkel (90), while (female) deputy leader Patricia Casale is not mentioned at all.21

Taken together, the website leaves no doubt that the AfD is a right-wing party. Only about 60 per cent of the references to “Germany” are due to the use of the full party name. The party simply talks a lot about Germany, everything German (253), Europe (379), the Euro (327), the EU (175), and the Eurozone (62). The tone is slightly harsher than that of the manifesto, with the occasional attack on the ECJ or refugees “who abuse the right to hospitality”. While “Volk” (people) is rare, the more intellectual “Bevölkerung” (population, 42) and particularly “Bürger” (citizen(s), 169) crop up much more often. Attacks on the AfD’s political competitors are also frequent: the CSU is mentioned 41 times, the FDP receives 75 references, and the CDU is mentioned 104 times.

But all in all, there is still little evidence of populism or right-wing radicalism. Immigrants and immigration are mentioned only 23 times (equivalent to a single mention in six per cent of all posts), and not necessarily in a negative context. Bulgaria and Romania are each referenced less than 15 times, Muslims hardly play a role at all, and even Turkey and the Turks are mentioned only 23 times. Remarkably, the AfD shows an unusual degree of sympathy for Russia and distrust for the US, both common in German Radical Right circles. But as far as foreigners are concerned, the main focus is clearly Greece and the Greeks (297 references).

The main website does not, however, include any interactive elements (guest books, comments, fora). Instead, the party relies on social media websites to interact with members, supporters, the media, and the general public.
Facebook is of particular importance for the party. As of July 2014, the official fanpage of the AfD’s federal organisation counts almost 122,000 “likes”.22 This is nearly twice as much as the SPD (just under 75,000) or the CDU (almost 84,000) can muster. On Twitter, the AfD federal organisation’s handle has only about 9,600 “followers”. The analysis will therefore focus on the AfD’s Facebook fanpage.

From a party’s point of view, fanpages are attractive because they provide a focus for political conversation about the party that is actually under the control of the party. Whereas communication on Twitter is largely unmoderated, spontaneous, and ephemeral, Facebook fanpages resemble traditional home pages. Crucially, fanpage administrators can remove posts, restrict who may post, and even ban individual users from the page.

Montage of Soundbite and Party Logo

Facebook has created an application programming interface (API) that makes it easy to programmatically access posts on fanpages as well as their meta data. Data were collected using version 0.4 of the “Rfacebook” package for R. The AfD launched its fanpage on March 3, 2013, and posted for the first time on March 7. As of July 11, 2014, the AfD have updated their status 1,702 times, or roughly 24 times per week, with markedly higher frequencies immediately after the launch of the party, during the federal campaign, during the candidate selection conference in January/February, and finally during the European parliamentary campaign.23 Many of these updates include images that combine text and pictures. Figure 2 is quite a typical specimen that brings together a photo of Gauland, a short quote (“Germany is not the USA’s doormat”), and the party logo and signature blue background. It is well known that such photo updates create quicker and stronger reactions with Facebook users and are privileged by Facebook’s selection algorithm. Moreover, text within images is never truncated by Facebook and can be easily shared both on Facebook and across other channels with minimal effort.

The use of such imagery shows the professionalism of the AfD’s social media team but is an obstacle for text analysis, because the use of various fonts and designs renders reliable optical character recognition virtually impossible. Fortunately, most images are complemented by some text, which usually re-iterates the main points or raises some additional issues. The following analysis is based on 1,223 posts that contain at least some text. Together, these posts make up some 72,000 words.24

Facebook is a popular medium for political communication because links to other content on the internet can be quickly posted (often with a preview of the other site’s content), distributed, and commented upon. Until July, the AfD had posted 1,622 unique URLs which point to 187 separate domains. From these figures, it is clear that the AfD does not simply use Facebook to re-publish the content of its main website. Indeed, there are only 69 links (less than five per cent) to the party’s national website, and 70 links to the websites of 19 local or regional party organisations.

The vast majority (795) of links refer to other content on Facebook. The party also makes extensive use of video clips hosted on youtube.com (79). Amongst the other sites, welt.de (the right-most mainstream broadsheet) and faz.net (a centre-right broadsheet) are particularly prominent with 109 and 60 links, respectively. Other important mainstream sources include the business news sites handelsblatt.com (68) and wiwo.de (29) as well as news magazines focus.de (92, centre-right) and spiegel.de (58, centre-left). Finally, there is a host of links to various blogs and other websites.

In summary, the AfD use their Facebook page to direct attention to news articles that support the party’s positions, to “spin” stories on issues that will chime with their supporters, and occasionally to poke fun at their political adversaries. But what exactly are they talking about in the text that accompanies links, videos, and images? Amongst the most frequent words in the posts are once more “AfD” (1,182), “Germany” (701), “Euro” (488) “alternative” (466), “EU” (400), and “Lucke” (367). Again, other party leaders are mentioned far less often. Greece (together with Greeks and “Athens”) feature prominently once more with 202 mentions and are far more important than Turkey/Turks (48), Muslims/Islam (20), or Romania (9) and Bulgaria (8).

Apart from that, it is slightly easier to find populist rhetoric on the Facebook page than on the main website. While “elites” (which by any reasonable definition would include many AfD leaders) are hardly ever mentioned, there are ample references to a conflict between “politicians” and “citizens” as well as many calls for protecting “freedom” and “democracy”. But even the most overtly populist post, the party anthem “We don’t give up” is relatively tame: Germans are a “really super people” (“ein wirklich tolles Volk”) who nonetheless “suffer”, Chancellor Merkel is accused of treating “us” like a bunch of “right-less monkeys” while politicians more generally are guilty of writing incomprehensible and self-serving laws. The only solution to this crisis is to vote for the AfD.

Collectively, the AfD’s posts were “shared” (copied to users’ profiles or other pages) more than 500,000 times. They received more than 1.9 million “likes” (on average, more than 1,100 per post) and over 325,000 comments, which amounts to over 200 comments per post.

Bloggers and mainstream journalists have repeatedly suggested that the AfD buys phantom fans and fake likes on Facebook. Comments, however, are much more difficult to simulate than shares and likes, and even a cursory glance at the AfD’s page shows a remarkable level of real political interaction between users.

Somewhat surprisingly, not just “fans” but any Facebook user may comment and even post new content on the AfD’s page. As of July 11 2014, almost 79,000 user-generated posts are accessible on the page. This is roughly equivalent to a corpus of 3.4 million words. Together, the posts have attracted more than 212,000 comments and just over 51,000 shares.

While members and supporters dominate, some critical voices exist amongst those who post on the AfD’s page. At 7,980, the number of original posters is relatively low, and the distribution of posts across users is heavily skewed to the right (18.7). The median number of posts per user is just one. A minority of five per cent has posted 30 times or more, and a tiny group of 25 users (less than 0.5 per cent) is collectively responsible for a quarter of all posts.25 While some of these 25 show off their sympathy for the AfD in their profile pictures and three hold party offices at the local level, none of them plays any significant role within the party or holds public office.

In terms of content, the 79,000 user-generated posts resemble the material posted by the AfD themselves. Again, “AfD” (26,311), “Germany” (12,816), “Euro” (11,152), and EU (9,295) are amongst the most frequent words, while “Bulgaria”, “Romania”, “Turkey”, and Muslims/Islam are of lesser importance. However, quite a few posts strike a tone that is markedly different from the party’s carefully crafted statements. Resentment and nationalism colour many posts. Complaints about ungrateful immigrants, privileged homosexuals, and greedy politicians are frequent. Links to obscure right-wing sites abound.

The AfD have created a space for their supporters where this kind of talk is tolerated. But only up to a point: Racist slurs and even common expletives are very rare. This does not prove, but suggests, continuous interventions by the party. In various comments, the AfD has made it clear that they are actively monitoring the page, and that they delete racist or otherwise illegal content including links to right-wing extremist websites. There is no way of knowing how many items have been posted and subsequently deleted, but the party is treading a narrow line. On the one hand, the AfD does not want to annoy their most vocal supporters on the internet, on the other, Lucke is very wary of allegations of populism and radicalism.

Conclusion

This article set out to answer the question whether the AfD is a right-wing, populist, and eurosceptic party. A careful quantitative and qualitative analysis of the 2014 EP manifesto shows that the AfD is indeed located at the far-right end of Germany’s political spectrum because of their nationalism, their resistance against state support for sexual diversity and gender mainstreaming, and their market liberalism. They do, however, not qualify as “radical”: There is no evidence of nativism or populism in their manifesto, which sets them apart from most of the other new right parties in Europe. Moreover, their euroscepticisim is of the “soft” variety. This assessment is largely confirmed by an analysis of the AfD’s communication on the web, although statements by their facebook fans hint at more radical currents amongst supporters and the party rank-and-file.

Important nuances not withstanding, their current programmatic appeal most closely resembles that of the CSU. But while the CSU is essentially an “ethno-regional” (Falkenhagen 2013) party that does not stand candidates outside Bavaria, the AfD aims at attracting a much bigger and broader national constituency.

Continued electoral support for the AfD would have profound repercussions for the existing German party system, most obviously by undermining the position of the CDU, which so far have fared much better than other Christian Democratic parties in Western Europe (Bale and Krouwel 2013). In the longer run, it would directly or indirectly affect domestic, immigration, and European integration policies. Thus far, Chancellor Merkel has ruled out coalitions with the AfD. If the FDP’s decline proves permanent and the AfD prevails, maintaining this cordon sanitaire will weaken the position of the CDU by forcing it to exclusively enter coalitions with the parties of the Left. At the same time, the AfD’s success is already fuelling internal backlash from conservatives against Merkel’s socially liberal policies.

Irrespective of the AfD’s perspectives for long-term survival, the fact that the party has done so well in five consecutive elections reflects the scope of partisan dealignment in Germany and the increasingly fluid nature of its party system. But does this remarkable mobilisation guarantee a bright future for the AfD? Not necessarily. In absolute terms, the AfD’s support has essentially stagnated, although the timing of the elections was close to optimal: In 2013, the AfD won 2.06 million votes, in the 2014 election it was 2.07 millions. In the eastern state elections, the AfD even suffered a small net loss of some 18,000 votes compared to their state-level results in the European election. Euroscepticism, the party’s current core issue, is still not very salient in Germany. In a survey immediately before the EP election, only 47 per cent of the AfD’s own voters had a wholly negative view of Germany’s membership in the EU. Only 22 per cent rejected both Juncker and Schulz as president of the commission, and 40 per cent said they supported the AfD to register a protest vote, not because of their policies.

Precisely what these policies are might therefore change over the near future. So far, the “moderates” have dominated the party leadership. Lucke has been able to commit the AfD to civic nationalism, financial prudence, and soft euroscepticisim, with Gauland and Adam catering for a more broadly “liberal-conservative” right-wing audience that may feel left behind by Merkel’s move to the centre. In 2014, Lucke’s creation more closely resembles the British Conservatives than UKIP, let alone the French FN or the Austrian Freedom Party.

Yet Lucke’s control over the party seems to be wavering. His plans to replace the joint leadership structure with a more traditional sole-leader role have been rejected by a party conference and have been met with criticism from his colleagues.26 State and district party chapters are still struggling to keep right-wing extremists out. Meanwhile, their new status as MEPs has given other leading figures such as von Storch and Pretzell a platform. They represent less savoury brands of right-wing politics that could ultimately prove more attractive to voters than Lucke’s polite exercises in economic theory. Just how long the party resists that temptation remains to be seen.

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1E.g. Stuttgarter Zeitung 18/09/2013, page 4; TAZ 02/07/2014, page 5.

2Most parties will of course try to alter this environment to their advantage.

3In a broader sense, the party’s organisational structures and deployment of campaign funds could also go under that rubric, although these obviously depend on recruitment and supply of external resources.

4The ‘Law and Order Party’ (PRO) of the early 2000s initially took a similar approach, but was not interested in euroscepticism and remained confined to the city-state of Hamburg.

5The original manifesto is archived at http://web.archive.org/web/20120923000310/http://www.wa2013.de/index.php?id=208 .

6Merkur Online, http://www.merkur-online.de/aktuelles/politik/alternative-deutschland-afd-zustrom-enorm-ueber-10000-mitglieder-zr-2873622.html (03/07/2014).

7RP Online 01/12/2013, http://www.rp-online.de/politik/deutschland/joerg-burger-ist-neuer-afd-chef-in-nrw-aid-1.3857120 (03/07/2014), Spiegel Online 28/12/2013, http://www.spiegel.de/politik/deutschland/afd-lucke-will-hessischen-landesvorsitzenden-abwaehlen-lassen-a-941072.html (03/07/2014), Zeit Online 14/06/2014, http://www.zeit.de/politik/deutschland/2014-06/afd-thueringen-ruecktritt (03/07/2014).

8Süddeutsche Online 23/03/2014, http://www.sueddeutsche.de/politik/europa-parteitag-afd-lehnt-sanktionen-gegen-russland-ab-1.1919526 (03/07/2014).

9Deutsche Welle Online 29/03/2014, http://www.dw.de/united-against-the-european-union/a-17530053 (03/07/2014).

10Lucke himself, Hans-Olaf Henkel, the former president of the Umbrella Organisation of German Industry (BDI), who favours a minimal state and has likened the EU to the former Soviet Union(Handelsblatt 03/10/2011, http://www.handelsblatt.com/meinung/kolumnen/kurz-und-schmerzhaft/henkel-trocken-use-eudssr/4681178.html (03/07/2014)), Bernd Kölmel, a public servant with the Baden-Württemberg State Court of Auditors, Beatrix von Storch, an insolvency lawyer and fringe Christian-conservative homphobe activist, Joachim Starbatty, a retired professor of economics who has repeatedly (though unsuccessfully) sued the government over the Euro, Ulrike Trebesius, a civil engineer, and the aforementioned Marcus Pretzell, a lawyer and property developer.

11The Graphs show the average of the boot-strapped point estimates. For the parameters that determine word “loadings” on the ideological dimension, these tend to differ somewhat from the maximum likelihood estimates, but for the parameters, the two sets are virtually identical.

12For the sake of readability, words are used here instead of the actual stems.

13However, in the vast majority of the 500 bootstrap samples, the CSU is estimated to be slightly more right-wing than the AfD.

14E.g. Spiegel Online 22/03/2014, http://www.spiegel.de/politik/deutschland/afd-parteitag-in-erfurt-bernd-lucke-attackiert-medien-a-960230.html (07/07/2014).

15Zeit Online 18/03/2013, http://www.zeit.de/2013/17/alternative-fuer-deutschland-ausrichtung (07/07/2014).

16With the tacit blessing of the leadership, a group of backbenchers have voted against the bailout legistlation and subsequently asked the Federal Constitutional Court to nullify these bills. The party’s core political project in the 2013-2017 parliament is a special road charge for cars registered abroad that would probably violate of EU law, and the central plank of their 2014 campaign was the slogan “kick welfare cheats out”, which referred to alleged “benefit tourists” from the eastern EU member states.

17This interpretation was emphasised by the design of the campaign posters, which surrounded the “EU” in “Deutschland” with the 12 European stars (shown in Figure 2).

18http://www.alternativefuer.de/konrad-adam-wie-wird-man-zu-einem-gefaehrlichen-buerger/ .

19More generally, Lucke and his party are the object of very intense scrutiny by their political adversaries and the mainstream media, yet there are very few verifiable public statements by Lucke or other members of the national leadership that could qualify as right-wing populist (see the borderline examples in Häusler and Roeser 2014, 37).

20This number excludes automatically generated overview pages for authors, tags, categories, and year of publication.

21Another relatively prominent male is Starbatty (20), now a MEP. On the other hand, there are only five references to female MEP von Storch , and the second female MEP Trebesius is mentioned just once. The image of the party leadership that the website projects is reflected in the coverage by the German media: For the period from February 1 2013 to July 11 2014, LexisNexis lists 2096 news items mentioning both the AfD and Lucke and 622 items that refer to Henkel and the AfD. Adam, however is mentioned only 232 times, Gauland 231 times and Petry 185 times. von Storch is referenced 169 times, Starbatty 150 times, Trebesius 20 times, Pretzell four times, while there is just one article that mentions Casale.

22All 16 state level chapters as well as various regional and local chapters have set up their own fanpages, but those have much smaller fanbases, ranging from several thousands to less than a hundred.

23This number could be inflated, as the Facebook API returns about 264 posts which do neither contain messages nor links, and which created no reactions. Presumably, these are either drafts or were retracted.

24This number includes URLs and symbols such as the hashtag sign.

25The number could be even smaller, as three of the most prolific posters have very similar surnames: Otto Blank, Andrea Blanc, and Andrea Cnalb.

26FAZ Online 20/10/2014, http://www.faz.net/aktuell/politik/machtkampf-in-der-afd-lucke-und-die-ruecktrittsdrohung-13220047.html (07/11/2014), Spiegel Online 29/11/2014, http://www.spiegel.de/politik/deutschland/afd-bernd-lucke-laut-alexander-gauland-ein-kontrollfreak-a-1003889.html (29/11/2014).

Rechtsextremismus in Deutschland in sechs handlichen Kapiteln [Rezensionsessay]

 

Gideon Botsch, Die extreme Rechte in der Bundesrepublik Deutschland von 1949 bis heute, Darmstadt 2012 (WBG), 151 S.

 Die extreme Rechte in der Bundesrepublik bildet zugleich ein kleines und ein sehr unübersichtliches Forschungsfeld: Klein, weil sich die Zahl der Akteure aufgrund des hohen Verfolgungsdrucks zumeist in überschaubaren Grenzen hielt, unübersichtlich, weil sich diese Akteure in einer Unzahl von teils sehr kurzlebigen Gruppen zusammengefunden haben und weil die Grenzen zur demokratischen Rechten zuweilen verschwimmen.

Gideon Botsch, Politikwissenschaftler und Mitarbeiter am Moses Mendelssohn Zentrum der Universität Potsdam, hat nun in der Reihe “Geschichte Kompakt” eine knappe Gesamtdarstellung der deutschen extremen Rechten nach 1945 vorgelegt, die einen komprimierten, aber sehr gut lesbaren Überblick über mehr als sechs Jahrzehnte bundesdeutschen Rechtsextremismus gibt. Naturgemäß läßt sich dies nur durch didaktische Reduktion und Verdichtung erreichen.

Botsch stellt seiner eigentlichen Arbeit eine Einleitung von 17 Seiten voran, in der er neben einer Diskussion von Grundbegriffen (Rechtsradikalismus, -extremismus, populismus, Antisemitismus etc.) und einem Überblick über den Aufbau des Buches noch eine Darstellung der extremen Rechten vom Kaiserreich bis zum Ende des Zweiten Weltkrieges unterbringt. Den Hauptteil des Buches bilden dann drei Kapitel, die unter den Überschriften “Nationale Opposition in der Nachkriegsgesellschaft”, “Nationale Opposition im Übergang” und “Nationale Opposition im geeinten Deutschland” jeweils zwei Dekaden in der Entwicklung der extremen Rechten in Deutschland behandeln, denen jeweils ein Unterabschnitt gewidmet ist. Den Zeitraum von 1980 bis 1989 beispielsweise untersucht Botsch unter der Überschrift “Zwischen Terror und Wahlkampf”. Innerhalb jedes Kapitels betrachtet der Autor vor allem Parteien und andere primär politisch ausgerichtete Organisationen. Die “Lebenswelt” der Jugend- und anderen Vorfeldorganisationen wird eher exemplarisch dargestellt. Eine sehr knappe Schlußbetrachtung, in der Botsch vor allem die aktuellen Entwicklungen bei der NPD und die Aufdeckung des “Nationalsozialistischen Untergrundes” kommentiert, schließt das Werk ab.

Botschs Periodisierung ließe sich trefflich kritisieren, da politische Ereignisse und Entwicklungen sich üblicherweise nicht an die Grenzen von Dekaden halten. Botsch selbst weist in diesem Zusammenhang auf die von Richard Stöss vorgeschlagene und in der Wahl- und Parteienforschung weithin akzeptierte alternative Einteilung in drei bzw. vier “Wellen” der rechtsextremen Wahlerfolge hin. Auch die politischen Biographien der (oft erstaunlich langlebigen) Akteure und der von ihnen begründeten Organisationen fügen sich selten nahtlos in das Dekadenschema ein.

Dennoch erscheint die von Botsch gewählte Einteilung in sechs Jahrzehnte, in denen die Entwicklung der extremen Rechten jeweils einer Art Leitmotiv folgt, erstaunlich plausibel. Dies liegt zum einen daran, dass einige für die extreme Rechte wichtige Ereignisse – das Verbot der SRP 1952, das Scheitern der NPD bei der Bundestagswahl 1969, der Anschlag auf das Oktoberfest 1980 und der Verbotsantrag gegen die NPD im Januar 2001 – tatsächlich auf halbwegs runde Jahrezahlen fallen und die Entwicklung in den darauffolgenden Jahren geprägt haben. Zum anderen aber geht es Botsch bei seiner Einteilung hauptsächlich um eine Didaktisierung des Stoffes. Für ihn stehen die Dekaden “jeweils als Begriffe für sich selbst und erzeugen unmittelbar eine Reihe von assoziationen und Bildern” (S. 6), die der Autor nutzbar macht, um seinen Lesern die ansonsten doch sehr unübersichtliche Entwicklungsgeschichte des neueren deutschen Rechtsextremismus als vergleichsweise wohlgeordnetes Tableau präsentieren zu können.

Mit diesem Zugang ordnet sich Botsch in das Programm der Reihe “Geschichte Kompakt” und einer wachsenden Zahl ähnlicher Projekte ein, die allesamt darauf abzielen, Lehrstoff für Studierende in den “neuen” Studiengängen in besonderer Weise aufzubereiten – die Buchgesellschaft ist sich nicht einmal zu schade dazu, den Umschlag mit einem “Bachelor/Master geprüft-Siegel” zu versehen, das den Eindruck erwecken soll, dass auch dieses Buch in irgendeiner Form “akkrediert” worden sei. Dagegen ist per se nichts einzuwenden. Im Gegenteil: Von Zeittafeln, besonders hervorgehobenen Definitionen, ausgewählten Quellen, Minibiographien und Randspalten profitieren alle Leser, die sich in dem Bändchen rasch zurecht finden wollen. Zu kritisieren ist hier allenfalls das Fehlen der auf dem Umschlag annoncierten “klar strukturierten Grafiken”, mit denen man den teils sehr verschlungenen Stammbaum der Rechtaußen-Parteien hätte illustrieren können.

Sehr bedauerlich ist aber das durch das Format erzwungene fast vollständige Fehlen von Fußnoten und der weitgehende Verzicht auf Belege im Text, die nur für einige zentrale wörtliche Zitate angegeben sind. Dadurch verliert Botschs Kompendium für Fachkollegen erheblich an Nutzen und vermittelt den Studierende ein – gelinde gesagt – irreführendes Bild von der wissenschaftlichen Methode. Dies ist schon deshalb problematisch, weil viele der von Botsch beschriebenen Gruppierungen klandestin organisiert sind. Für Studierende und Doktoranden wäre es wichtig zu wissen, auf welcher Grundlage ein so schwieriges Feld überhaupt erforscht werden kann.

Auch die Auswahlbibliographie ist mit zweieinhalb Seiten deutlich zu kurz geraten. Hier fehlt es – gerade vor dem Hintergrund der unzureichenden Belege im Text – nicht nur an Quellen, sondern vor allen Dingen an Kommentaren, die diese erschließen und den tatsächlich interessierten Studierenden Hinweise geben, wie sie sich Teilaspekte des Themas selbst erarbeiten können.

Ein ganz erhebliches weiteres Problem liegt darin, dass die Darstellung fast vollständig ohne einen theoretischen Unterbau auskommen muß und sich statt dessen primär an der eingangs erwähnten Einteilung in Dekaden sowie der (nicht näher erläuterten) Idee von “Ereignisketten” orientiert. Eine politikwissenschaftlicher oder extremismustheoretischer Rahmen fehlt. Konzepte wie das “Angebot” von und die “Nachfrage” nach extremistischen Politikinhalten werden zwar in der Einleitung erwähnt, spielen aber für das Folgende kaum noch eine Rolle. Die eingestreuten Erklärungen für die Entwicklungen innerhalb der extremen Rechten bzw. der bundesdeutschen Gesellschaft insgesamt sind deshalb zwar durchaus plausibel. Eine Einbettung in einen größeren Argumentationszusammenhang ist aber kaum erkennbar. Dies ist bedauerlich, weil dadurch trotz der faktengesättigten Darstellung einige der interessantesten Fragen der Rechtsextremismusforschung gar nicht erst in den Blick geraten.

So hat beispielsweise Herbert Kitschelt bereits vor fast 20 Jahren die Hypothese aufgestellt, dass die extreme Rechte in Deutschland regelmäßig an ihrer Fixierung auf und ihrer Verbindung zur jüngsten deutschen Geschichte scheitert und sich deshalb – anders als etwa in Frankreich, Skandinavien, der Schweiz und Österreich – bislang keine moderne und erfolgreiche Rechtsaußen-Partei etablieren konnte. Als Beleg für die Gültigkeit von Kitschelts Hypothese wird häufig die NPD genannt, die sich noch in den frühen 1980er Jahren vornehmlich mit den verlorenen Ostgebieten, Kriegsschuld- und Holocaust-Debatten und nicht zuletzt dem klassischen Antisemitismus beschäftigte und darüber die aufkommenden Fragen der Asyl- und Zuwanderungspolitik fast übersehen hat.

In seinen Ausführungen zu “Terror und Wahlkampf” beschreibt Botsch nun in wenigen dürren Sätzen die Entdeckung des Migrationsthemas und die halbherzigen und kurzlebigen Versuche der NPD, sich in einem neuen Programm von ihrer seit Jahrzehnten geführten vergangenheitspolitischen Debattenkultur zu lösen (S. 88-89), ohne die Signifikanz dieses Manövers deutlich zu machen oder auf die Gründe für sein Scheitern einzugehen. In ähnlicher Weise wird auch die Hinwendung der “Republikaner” zum Rechtsextremismus nach der Machtübernahme durch Franz Schönhuber lapidar als Faktum präsentiert, aber nicht in einen größere Kontext gestellt. Botschs Darstellung der Vorgänge ist an dieser wie an anderer Stelle konzise und hochinformativ, würde aber durch einen stärker analytischen Zugriff, durch ein Mehr an Interpretation, vielleicht sogar durch zielgerichtete Spekulation – was wäre gewesen, wenn sich die Republikaner bereits früher von der NS-Verehrung und anderen verfassungsfeindlichen Tendenzen in ihren Reihen überzeugend distanziert hätten – nochmals gewinnen. Generell wird das (insgesamt bislang ebenfalls recht erfolglose) rechtspopulistische Spektrum in Deutschland – zu nennen sind hier neben älteren Gruppierungen wie der “Partei Rechtsstaatliche Offensive”, “Pro-DM” und der “Bund freier Bürger” vor allem die “Pro-Bewegung” (die Botsch eher dem Rechtsextremismus zuordnet) und “Die Freiheit” – nur am Rande betrachtet. Auch zum Übergangsbereich zwischen etablierten Parteien und extremer Rechter und hier insbesondere zum gelegentlichen Aufflackern nationalliberaler bis nationalpopulistischer Strömungen in der FDP hätte man gerne noch etwas mehr gelesen.

Wie oben bereits angesprochen, betrachtet Botsch außer der parteipolitisch organisierten Rechten auch sehr intensiv die Lebenswelt des rechten Milieus, d.h. das Netzwerk, das von Zeitschriften und Tagungshäusern über “Jugendbünde” bis hin zu Kameradschaften und kaum organisierten Schlägerbanden reicht. Mit guten Gründen beschränkt sich Botsch hier auf eine exemplarische Darstellung, auch wenn die Auswahlkriterien, nach denen er vorgeht, nicht immer ganz klar werden.

Die relativ ausführliche Beschäftigung mit den “Bünden” und “Ringen” ist sicher deren historischer Bedeutung insgesamt sowie ihrer früheren Rolle als zentrale Sozialistions- und Rekrutierungsinstitutionen geschuldet, die sie aber seit den 1970er Jahren weitgehend verloren haben dürften. Für den Zeitraum seit den 1980er Jahren ergibt sich aus Botschs Darstellung (z.B. auf S. 120-121) der Eindruck, dass viele Organisationen im Grunde nur noch von und für einige wenige Familien weitergeführt werden, die sich seit Generationen dem Nationalsozialismus verschrieben haben.

Angaben zur Zahl der Mitglieder dieser Gruppierungen sind sicher schwer zu ermitteln, grobe Schätzungen wären aber hilfreich. Umgekehrt hätte man als Leser gerne mehr über die sogenannten “Freie Kameradschaften”, über sonstige Neonazigruppen und vor allem über die rechte Subkultur im deutschsprachigen Internet gewußt, die für viele Aktivisten und Sympathisanten einen weitgehend rechts- und repressionsfreien Raum bildet, der traditionelle Formen der rechten Jugendkultur zumindest ergänzt, wenn nicht sogar partiell ersetzt.

Sehr informativ wenn auch knapp sind schließlich Botschs Ausführungen zum Rechtsterrorismus insbesondere der 1980er Jahre. Zurecht weist der Autor hier darauf hin, dass dieser im Gegensatz zum Terror der RAF und der Roten Zellen von Medien und Öffentlichkeit weitgehend vergessen worden ist, obwohl die Zahl der Opfer hoch war und einige der sogenannten “Wehrsportgruppen” über paramilitärische Ausrüstung und entsprechendes Training verfügten. Hier wäre – gerade im Zusammenhang mit den Morden durch den sogenannten “Nationalsozialistischen Untergrund” und den spektakulären Fehlleistungen der Sicherheitsbehörden bei deren Aufdeckung – noch intensiver nach den strukturellen Ursachen und Folgen dieser Fehlwahrnehmung zu fragen.

Abschließend stellt sich die Frage, an welches Publikum sich Botschs Werk richtet. Für Fachwissenschaftler hat es trotz fehlender Fußnoten und sonstiger Belege einen gewissen Nutzen als komprimierter Überblick über die wichtigsten Stationen des Nachkriegs-Rechtsextremismus. Auch für die eigentliche Zielgruppe, d.h. für Studierende in den BA- und MA-Studiengängen ist es im Grunde gut geeignet, sollte bei der Kursplanung aber nicht als Lehrbuch, sondern vielmehr als ein Nachschlagewerk betrachtet werden, mit dessen Hilfe sich die Studierenden rasch das notwendige zeitgeschichtliche Hintergrundwissen erschließen können. Für eine ernsthafte Auseinandersetzung mit der extremen Rechten bleiben aber auch in Zeiten von Bachelor und Master ein solides theoretisches Rüstzeug und die Auseinandersetzung mit der aktuellen Forschungsliteratur unabdingbar.

Political Opportunity Structures and Right-Wing Extremist Party Success

 

West European right-wing extremist parties have received a great deal attention in the academic literature due to the success that many of these actors have experienced at the polls. What has received less coverage, however, is the fact that these parties have not enjoyed a consistent level of electoral support in this third wave of right-wing extremist party activity (Beyme, 1988). Instead, their electoral fortunes have risen and fallen over the last two decades. The fact that this question of variation in the electoral support for the parties of the extreme right – both over time and across countries – has attracted relatively little attention in the literature is not overly surprising. For one thing, there continues to be a shortage of comparative studies on the extreme right and in particular on the extreme right’s voters. In addition, as far as the studies that do exist are concerned, it is not surprising that many of these have tended to focus only on why right-wing extremist parties have been successful, rather than on why they have not.

 

The few works that have addressed the issue of the variation in the electoral support for the parties of the extreme right across Western Europe have tended to offer only partial explanations for this phenomenon. Jackman and Volpert (1996), for example, assess the importance of electoral system, party system and economic factors on the right-wing extremist party vote, but they do not consider the impact of different socio-demographic variables. Likewise, Abedi (2002) concentrates on the effect of party system factors but fails to examine the influence of socio-economic variables and of other institutional characteristics. Knigge (1998), by contrast, explores the effect of some socio-economic factors but does not examine the impact of electoral system or party system factors. Thus, while these studies each add to an overall explanation for the variation in the electoral fortunes of the parties of the extreme right, on their own, they offer an account for the phenomenon that is far from comprehensive.

 

A more extensive explanation for the uneven electoral success of the parties of the extreme right is to be found in the influential work by Kitschelt (1995) and in the useful study by Lubbers and his colleagues (2002). However, in spite of its comprehensive nature and of the significant contribution that it makes to research on right-wing extremism, the study by Kitschelt also has a number of limitations. In particular, the framework employed does not allow for a precise assessment of the relative influence of the different independent variables on each of the right-wing extremist parties under observation.

 

The study by Lubbers et al. certainly does not suffer from this limitation. Yet, it is problematic, too, in terms of its methodology, the countries that it covers and its time-span. The decision to combine data from national election studies with data sets from supra-national projects raises potential problems of validity and reliability. In addition, the use of multi-level analysis is open to question.i As for the countries examined, the inclusion of countries where support for the extreme right is extremely low is also not without consequences. Finally, in terms of the time-span covered, Lubbers and his colleagues analyze data from 1994 to 1997 only, and do not cover the early to mid-1980s in which many right-wing extremist parties of the third wave broke through into the electoral arena. Therefore, the variance in explanatory factors such as unemployment, immigration and the positions of other parties is probably severely restricted.

 

In light of the limitations of the existing studies, this paper seeks to put forward an explanation for the variation in the right-wing extremist party vote across Western Europe that incorporates a wider range of factors than have been previously considered and that covers a longer time period. More specifically, through the construction of an individual-level model, the paper first examines the impact of socio-demographic variables on the right-wing extremist party vote. Then, by augmenting the model with system-level information, the paper investigates the influence of a whole host of structural factors (which together make up the political opportunity structure) that may potentially affect the extreme right’s performance at the polls. This two-stage approach enables us to assess the extent to which system-level features (relating to the political opportunity structure) account for variation in the extreme right’s success over time and across countries after individual-level features have been controlled for. Moreover, it also allows us to establish whether individual-level characteristics still have an effect on right-wing extremist voting when the political opportunity structure is held constant. The paper concludes with an assessment of which variables have the most power in explaining the uneven electoral success of right-wing extremist parties across Western Europe.

 

 

Theoretical Framework

 

i) Socio-demographic Factors

 

It has been well documented that certain socio-demographic groups have shown themselves more likely to vote for the parties of the extreme right than others. In the first instance, a significant gender gap in the support for the extreme right has been reported, with male voters exhibiting a greater propensity to vote for right-wing extremist parties than their female counterparts (see for example Betz, 1994; Lubbers et al., 2002).

 

Similarly, the existing studies have shown that an age effect exists, with both younger and older voters being more likely to support the extreme right than other age groups. A number of theories help explain this U-shaped phenomenon. It has been well documented, for example, that the decline in the effects of social structure has not affected all generations equally, and that younger voters as well as pensioners are more likely to lack social ties. Greater social integration is likely to be reflected not only in higher levels of electoral participation but also in a tendency to refrain from voting for a party of the extreme right. A further explanation for the greater propensity of both young and older voters to support the extreme right rests in these people’s interests and their access to welfare. Since young and old voters depend disproportionately on welfare, these two age groups are more likely to view immigrants as competitors than are people of other age groups.

 

As regards formal education, it is often hypothesized that people with lower levels of education will exhibit a greater propensity to vote for parties of the extreme right than people with higher levels of education. In the first instance, there is an economic or an interest-based argument to support this presumption: voters with lower levels of education tend to be less skilled, and hence are likely to fall victim to market forces (Falter, 1994: 69). They tend to support parties of the extreme right because these parties pledge to defend the economic interests of these voters by limiting the rights of immigrants and asylum-seekers, who are perceived as direct competitors both in the workplace and in accessing social services and housing. Another argument is value-based. It rests on the premise that, through education, people are intensively exposed to liberal values, and hence the longer a person spends in education, the more likely they are to embrace such values (Warwick, 1998; Weakliem, 2002). A similar argument holds that cognitive style effects explain the link between a person’s propensity to vote for a party of the extreme right and their level of education (Weil, 1985).

 

Finally, as regards class, a number of national studies have shown shopkeepers, artisans and small-business people to be particularly well represented among the electorates of right-wing extremist parties in several countries. An over-representation of working-class voters among those who support the parties of the extreme right – in some cases right from the start, in other instances growing over the years – is also well-documented by many studies at the national level. Finally, it has also been argued that people in non-manual jobs who enjoy a small degree of autonomy in their work may also develop authoritarian preferences, quite similar to those ascribed to working-class voters (Kitschelt, 1994: 16-17).

 

To sum up then, based on the evidence that has emerged in much of the existing literature, we expect there to be a greater propensity to vote for parties of the extreme right among men, among voters who are either young or old, among those with lower levels of formal education, and among the working class, the self-employed and those in routine non-manual forms of employment as compared to all other socio-demographic categories of elector

 

 

ii) Political Opportunity Structures

 

To assess the influence of structural or environmental factors on the right-wing extremist vote we draw on the concept of political opportunity structures, which was originally developed in the context of research on social movements to denote the degree of ‘openness’ or ‘accessibility’ of a given political system for would-be political entrepreneurs. In a very influential study Kitschelt describes political opportunity structures as ‘specific configurations of resources, institutional arrangements and historical precedents for social mobilization, which facilitate the development of protest movements in some instances and constrain them in others’ (1986: 58). As their name implies, political opportunity structures therefore emphasize the exogenous conditions for party success and, in so doing, contrast to actor-centred theories of success (Tarrow, 1998: 18).

 

The concept of political opportunity structures is a broad one and different authors have included different items in their definition of the term. In spite of the differences, however, the majority of studies agree that fixed or permanent institutional features combine with more short-term, volatile or conjectural factors to produce an overall particular opportunity structure (e.g. Kriesi et al., 1995). We therefore propose to adopt a three-pronged approach with which to examine the influence of political opportunity structures on the right-wing extremist party vote: a first set of variables captures the impact of long-term institutional features on the parties of the extreme right; a second set examines medium-term factors which relate to the party system; and a third set of variables examines short-term contextual or conjectural variables.

 

a) Long-term Institutional Variables

 

Two institutional variables we regard as being of potential importance to how well parties of the extreme right perform at the polls are (i) the electoral system, and (ii) the degree of decentralization/federalism. As far as electoral systems are concerned, it has long been established that the more proportional the electoral system, the greater the incentives for political entrepreneurs to enter the electoral race and for voters to decide to support a new or a small political party. By contrast, the less proportional the electoral system, the more leaders of new or small parties will be dissuaded from fielding candidates and the more discouraged voters will be from voting for such parties since they stand little change of gaining representation (Duverger, 1951; Blais and Carty, 1991). In view of this relationship, we anticipate that unless they have already reached a certain size and have a chance of continuing to attract a sizable section of the electorate, right-wing extremist parties are likely to suffer from disproportional electoral systems.

 

The effect of decentralization or federalism is less clear-cut. On the one hand, it can be argued that a high degree of decentralization (including regional parliaments) may foster the development of right-wing extremist parties because voters are more willing to support new and/or radical parties in ‘second order’ elections (Reif and Schmitt, 1980). However, rather than allowing extremist parties to gain a toehold in the electoral arena, it may instead be the case that second order elections serve as a kind of security valve for the political system by providing citizens with an opportunity to express their political frustration with the mainstream parties without overly disturbing the political process on the national level. Therefore, two contrasting – yet equally convincing – hypotheses as to the effect of territorial decentralization exist.

 

b) Medium-term Party System Variables

 

Party system variables are less constant than institutional factors. For reasons of parsimony, we restrict ourselves to examining the impact of three such variables: (i) the ideological position of other competitors in the party system, (ii) the degree of convergence between the mainstream parties, and (iii) the coalition format in the respective party systems.ii

 

We expect the position of the major party of the mainstream right in each of the respective party systems to have an impact on the success of the party of the extreme right, yet it is difficult to predict the exact nature of this impact. On the one hand it can be argued that the more right wing the party of the mainstream right, the less political space will be available to the party of the extreme right. On the other hand, it can be argued that a more right wing party of the mainstream right might legitimize the issues around which the extreme right mobilizes. Thus, two competing hypotheses emerge as to the influence of the ideological position of the mainstream right on the electoral success of the extreme right.

 

Next, we examine the degree of convergence between the parties of the mainstream right and the parties of the mainstream left in each of the party systems under observation.iii Here too, two contrasting hypotheses present themselves. On the one hand we can argue that right-wing extremist political parties will benefit where the mainstream right and the mainstream left converge (Kitschelt, 1995: 17). In such instances the parties of the extreme right can credibly argue that if voters wish to see a real alternative to both the government and the mainstream opposition, then they should put their support behind the right-wing extremist party. When the mainstream parties are ideologically distinct from each other, it is more difficult for the parties of the extreme right to adopt this strategy. On the other hand, the extreme right might perform well at the polls when the mainstream parties are ideologically quite distinct. First, this distinctiveness may signal the lack of elite consensus (Zaller, 1992), which might further extreme right party success. Second, the mainstream parties may have diverged ideologically in an attempt to curb the advance of the parties of the extreme right in upcoming elections. Either way, ideological divergence between the mainstream parties may be associated with extreme right party success. Once again, therefore, two conflicting hypotheses exist as to the effect of ideological convergence of the mainstream parties on the right-wing extremist party vote.

 

We then move to consider the coalition format of the party systems under investigation. We suspect that the extreme right will benefit from grand coalitions because (i) voters will feel that there is a lack of other political alternatives during a grand coalition and (ii) supporters of the mainstream right may become alienated if they do not see their preferred policies being enacted and do not enjoy the consolation of seeing their party play the role of a principled opposition (Kitschelt, 1995: 17). Therefore, we anticipate that the right-wing extremist party vote will be higher in (or shortly after) periods of grand coalition government than it will be in periods of alternating government.

 

c) Short-term Contextual Variables

 

In addition to being affected by long-term institutional variables and medium-term party system variables, it is also reasonable to expect the right-wing extremist vote to be influenced by a number of short-term contextual factors. More specifically, given the considerable emphasis parties of the extreme right place on the issue of immigration from non-EU countries and on the supposed competition between immigrants and the indigenous population, we anticipate that levels of immigration and unemployment (both straightforward levels and also change in these levels) will exert an effect on how well the parties of the extreme right perform at the polls. We expect the right-wing extremist vote to be positively correlated to both the level of immigration and the level of unemployment.

 

 

Data and Methodology

 

The data in our analysis come from national election studies. The pooling and harmonizing was carried out under the auspices of the Extreme Right Electorates and Party Success (EREPS) Research Group.iv The major advantage of using national election studies is that they reflect voter behaviour at election time. This contrasts to supranational surveys, which may be carried out at a time close to the beginning of the electoral cycle in one country, but near the end of the cycle in another.

These national election studies provided us with information on the individual vote choices and the socio-demographic characteristics of West European electors. In contrast to some of existing studies of right-wing extremist electorates (e.g. van der Brug et al., 2000; Swyngedouw, 2001; Lubbers et al., 2002; van der Brug and Fennema, 2003), we do not include variables that capture the different attitudes of voters because there are very substantial problems in finding comparative indicators of attitudes in national election studies, both over time and across countries. Although there is clearly some trade-off to be had in deciding not to include attitudinal variables, we believe that the advantages of using national elections studies (rather than supranational surveys) outweigh any disadvantages that result from excluding attitudinal variables. Furthermore, in contrast to attitudinal data, socio-demographic data are relatively easily compared and are measured with much less error.

 

The countries included in our analysis are: Austria, Belgium,v Denmark, France, Germany Italy and Norway.vi This means that the parties included in our analysis are: the Freiheitliche Partei Österreichs (FPÖ), the Vlaams Blok (VB); the Fremskridtspartiet (FRPd) and the Dansk Folkeparti (DF); the Front National (FN); the Deutsche Volksunion (DVU), the Nationaldemokratische Partei Deutschlands (NPD) and the Republikaner (REP); the Movimento Sociale Italiano / Alleanza Nazionale (MSI / AN) until 1995;vii and the Fremskrittspartiet (FRPn).

 

In contrast to the study by Lubbers et al., we have excluded countries where support for the extreme right is extremely low. While we recognize that including countries in which there is no effective extreme right is certainly necessary in a macro-level explanation of the extreme right’s success (and failure to do so would result in selection bias), we believe that incorporating such countries in an analysis of individual voting decisions is problematic for three reasons: (i) voting for the reasonably established extreme right parties in countries like Belgium, France or even Germany is not comparable to voting for a tiny (and often fanatical) political sect, (ii) in countries like Portugal, Spain, Great Britain, and Ireland, extreme right voters are extremely rare, with their numbers in social surveys even lower than the electoral results suggest,viii and (iii) in countries where the extreme right is very weak, prospective extreme right voters are often prevented from supporting an extreme right party because candidates of these parties are only fielded in certain constituencies. which is not reflected in surveys, as such voters are coded either as non-voters or as supporters of another party. Therefore, the inclusion of survey data from countries where support for the extreme right is extremely low or non-existent therefore dilutes and distorts any analysis of individual voting decisions.

 

While the parties included in our analysis differ from each other in terms of their precise ideological profile, we nonetheless believe that they belong to the same party family, and that they can thus be treated as constituent members of a larger, single group. There has been much debate in the literature over the exact definition of right-wing extremism, and hence over which parties belong to the extreme right party family, but a consensus has nonetheless emerged within this body of work that a separate extreme right party family does indeed exist. While it is perhaps more heterogeneous than other party families, its constituent parts are distinct from the parties of the mainstream right, and they also share a number of ideological features (in particular some combination of racism, xenophobia, nationalism, and a desire for a strong state and law and order), which allow them to be grouped together at the far right end of the left-right political spectrum (see Ignazi, 1992; 2003; Hainsworth, 1992, 2000; Betz, 1994; Mudde, 1996, 2000 among others for further details of this debate). Further evidence of the fact that the parties included in our study belong to a common extreme right party family can be found in the series of expert judgments studies that have been carried out since the beginning of the 1980s (Castles and Mair, 1984; Laver and Hunt, 1992; Huber and Inglehart, 1995; Lubbers, 2000).

 

Our timeframe spans the years 1984-2001. Our start date is informed by the broad consensus in the literature on right-wing extremism that the 1980s saw the beginning of a third wave of right-wing extremist activity in Western Europe (Beyme, 1988). The majority of scholars of right-wing extremism also agree that the Scandinavian Progress Parties only became part of the right-wing extremist party family in the mid-1980s when refugee and immigration policies became their primary concerns (Kitschelt, 1995: 121; Goul Andersen and Bjørklund, 2000: 203-204; Hainsworth, 2000). We therefore began with the Danish election survey of 1984, and collected all available data for polities where the extreme right was a relevant player in national parliamentary elections.

 

The socio-demographic variables included in our model are the standard ones: gender, age (up to 24 years, 25-34 years, 35-44 years, 45-54 years, 55-64 years, 65 years and older), formal education (no education/primary education, mid-school, secondary education, university degree), and social class (measured by a simplified Goldthorpe classification: professionals / managers, routine non-manual, self-employed, manual).

 

For our examination of the influence of political opportunity structures on the right-wing extremist party vote, we augmented the socio-economic data derived from the national election studies with information on the political systems and the party systems of the countries under investigation. To assess the impact of institutional variables we made use of data (derived from Carter, 2002) that measured the disproportionality of the electoral systems according to the Gallagher index (Gallagher, 1991), and we adopted Lijphart’s index of federalism to reflect the degree of territorial decentralization (Lijphart, 1999). This ranges from 1 to 5, with 1 indicating a unitary and centralized state and 5 referring to a federal and decentralized state.

 

To explore the influence of the position of other political competitors, and to assess the impact of mainstream party convergence and divergence we drew on the data of the Comparative Manifesto Project (CMP) (Budge et al., 2001). From the CMP data we constructed a measure based on the parties’ policies on the issues of multiculturalism, internationalism, the ‘national way of life’, and law and order. While reflecting many of the components that make up the overarching left-right dimension, these policy items are particularly important to the parties of the extreme right as it is primarily along these dimensions that they compete with their mainstream rivals.ix Like all measures that are based on CMP data, it reflects the balance between ‘left’ and ‘right’ statements of a party. Negative figures indicate a leaning to the left, with an empirical minimum of -12.4 for the Norwegian Socialists in the 1990s, while positive numbers indicate a leaning to the right of this dimension. Here, the empirical maximum for a party that is considered a part of the moderate right is 20.4, achieved by the Danish KF during the 1990s. However, the major right parties usually register much lower scores like e.g. 3.6 for the Austrian ÖVP in 1994 or 3.3 for the French RPR in 1997.

 

To examine the effect of a grand coalition in the period directly before a general election, we drew on data from EJPR data Yearbooks and included an appropriate dummy variable in our model.

 

Finally, to evaluate the effect of conjectural factors on the decision to vote for the extreme right, we drew on unemployment data at the aggregate levelx, and on data reflecting the number of asylum seekers in the countries under observation.xi We included a measure of the yearly number of asylum-seekers per thousand inhabitants,xii and a measure of the yearly percentage of unemployed people in the total workforce. We also included change rates for both variables in our model because, according to the classical ‘J-curve’ reasoning (see Davies, 1974; Coenders and Scheepers, 1998), people might respond to changes rather than to the actual level of both measures.

 

In terms of methodology, we estimate a logit model with contextual variables. Our model thus allows us to estimate the probability of a voter voting for a party of the extreme right conditional on (i) his/her individual socio-demographic attributes, and (ii) the particular political opportunity structures present in his/her country at the time of the election. Since there is no strong theoretical argument as to why socio-demographic or system-level explanations for an extreme right vote should vary systematically over countries and across time,xiii we assume that the true regression coefficients are constant across countries and across time after controlling for both individual and contextual variables. Therefore we refrain from inserting dummies and interactions to capture cross-country differences in intercepts and slopes.

 

 

Findings

 

Looking at Table 1, we can see that our findings are in line with much of the previous research in the field.xiv The results show that being male substantially raises the odds of voting for the extreme right. Put differently, depending on the respondent’s other attributes, being male increases the probability of an individual being an extreme right voter by more than 50 percent. This coefficient suggests that there is a substantial gender-gap in the support for the extreme right voting in Western Europe even when we control for other socio-demographic variables such as age, education, and social class.

 

Turning to the influence of age, Table 1 illustrates the U-shaped effect of this variable that we expected to see. Wald tests show that there are no significant (p = 0.45) differences in the respective levels of extreme right support among those voters who are between the ages of 35 and 64, while the level of support for parties of the extreme right among both younger and older voters is higher.xv The propensity to vote for a party of the extreme right among voters who are aged between 25 and 34 is identical (p = 0.91) to that of the reference group (voters who are 65 or older), while voters who are younger than 25 years old are much more likely to vote for the extreme right than any other voters, including the reference group.

 

 

Table 1: Socio-demographic model

 

 

Independent Variables

b

eb

Male

0.476**

1.609**

(0.036)

(0.059)

Age: -24

0.280*

1.324*

(0.124)

(0.165)

Age: 25-34

-0.012

0.988

(0.114)

(0.113)

Age: 35-44

-0.174

0.841

(0.095)

(0.080)

Age: 45-54

-0.223**

0.800**

(0.074)

(0.059)

Age: 55-64

-0.186

0.830

(0.112)

(0.093)

No/Primary Education

0.388

1.474

(0.304)

(0.448)

Mid-School

0.832**

2.299**

(0.244)

(0.560)

Secondary School

0.624**

1.866**

(0.147)

(0.273)

Professionals/Managers

-0.054

0.948
(0.338) (0.320)

Routine Non-Manual

0.116 1.123
(0.264) (0.296)

Self-employed

0.243 1.275
(0.252) (0.321)

Manual

0.345 1.412
(0.186) (0.262)

Constant

-3.239**
(0.235)

Observations

50 276

Adj. Pseudo-R2 (Mc-Fadden)

0.03

BIC

-515 293

 

Notes:

Robust standard errors adjusted for clustering on country are shown in parentheses (see note Error: Reference source not found).

* significant at 5%; ** significant at 1%

 

 

As regards levels of formal education, we predicted that people with lower levels of education would exhibit a greater propensity to vote for parties of the extreme right than people with higher levels of education. When we examine our model, however, things are not as clear-cut. While the low level of support that extreme right parties receive from university-educated voters (the reference group) is in line with the predications advanced above, the coefficient for the group of voters with no education or with primary education is smaller than expected and is not significantly different from zero. We find, instead, that it is people with ‘mid-school’ diplomas who appear to form the core social base of the extreme right. Depending on his or her other characteristics, having a mid-school education more than doubles the probability of an individual voting for the extreme right. The effect of being educated to secondary level is somewhat weaker, but the difference between the two coefficients is not significant (p=0.19).

 

As concerns the effect of class, our findings are generally in line with our expectations. The results show that professionals and unclassified voters (the reference group) exhibit the lowest propensity to support extreme right-wing parties while the odds of an extreme right vote are somewhat higher if the respondent has a routine non-manual job, if he or she is self-employed, or if he or she is a manual worker.xvi

 

In a bid to summarize our socio-demographic findings we calculated the expected probability of an extreme right vote across varying levels of the independent variables (see Table 2). For the sake of brevity, we restricted class to unclassified voters (the reference group) in the upper section of the table, and to workers (the group with the highest propensity to vote for a party of the extreme right) in the lower section of the table. Above all, Table 2 shows the significant variation in the support for the extreme right that exists across the different socio-demographic groups. If, for example, we compare the predicted probability of a vote for the extreme right being cast by a female voter, aged 24 or less, with a university education and whose class is ‘unclassifiable’ with the predicted probability of an extreme right vote being cast by a male voter from the same age group, with a mid-school education and a manual job, we can see the full extent of this variation. Indeed, the figures in Table 2 illustrate that the predicated probability of the female voter just described voting for a party of the extreme right is roughly 5 percent (as shown in bold in the upper section of the table), whereas the predicted probability of the male voter just described voting for the extreme right is roughly 21 percent (as shown in bold in the lower section of the table). This example clearly illustrates that gender and education in particular have a sizeable impact on the probability of a person voting for a party of the extreme right, while age and class are somewhat weaker predictors.

 

 

Table 2: Predicted probabilities (in percent) of an extreme right vote, depending on gender, age, education, and social class.

 

class: unclassified

Female

Male

Age/Educ

no/primary

mid

secondary

university

no/primary

mid

secondary

university

-24

7

11

9

5

11

16

13

8

25-34

5

8

7

4

8

13

10

6

35-44

5

7

6

3

7

11

9

5

45-54

4

7

6

3

7

10

9

5

55-64

5

7

6

3

7

11

9

5

65-

5

8

7

4

9

13

11

6

class: manual

Female

Male

Age/Educ

no/primary

mid

secondary

university

no/primary

mid

secondary

university

-24

10

14

12

7

15

21

18

11

25-34

7

11

9

5

11

17

14

8

35-44

6

10

8

4

10

15

12

7

45-54

6

9

8

4

10

14

12

7

55-64

6

10

8

4

10

15

12

7

65-

8

11

9

5

12

17

14

8

 

Notes:

Typical 95%-confidence intervals based on robust standard errors adjusted for clustering on country: female, less than 25 years old, university educated, class ‘unclassified’: 2.9 – 8.2;

male, less than 25 years old, mid-school education, manual worker: 13.2 – 32.6.

 

 

So far, therefore, our discussion has illustrated that a voter’s socio-demographic attributes go a long way in helping to explain his or her propensity to vote for a party of the extreme right at election time. In addition to this, our results have by and large also been in line with those of many of the existing studies on right-wing extremism. In particular, our comparative study of 24 elections in 7 countries confirms that parties of the extreme right are strongest among the more marginalized sections of society, and that (when we control for other socio-demographic variables) their support is predominantly male.

 

This agreement with the existing studies notwithstanding, our results point to another important finding: the low adjusted (McFadden) pseudo R2 in our model (a mere 0.03) indicates that the variation in the electoral success of right-wing extremist parties both over time and across space cannot simply be explained by the different composition of the respective electorates. Instead, the variation in the electoral fortunes of the parties of the extreme right must be explained by factors other than socio-demographic ones.

 

To confirm this we added a series of dummies for the 24 elections under study in our model (not shown) so as to create a model that captured all variation in the extreme right vote that could potentially be due to system-level factors. The resulting R2 of 0.09 was substantially higher than the R2 of the model in Table 1, thereby indicating that the extreme right’s electoral success varies considerably over time and across space even if we control for the composition of the electorate. In light of this, we now augment our socio-demographic model shown in Table 1 with variables that relate to the political opportunity structure as discussed above.

 

Table 3 shows the results of the full model. Looking at the table, the first observation to make is that the coefficients for the socio-demographic variables have not greatly changed since we have augmented the model with the political opportunity structure variables.xvii Second, we see that some of the additional variables have statistically significant and sizeable effects on an individual’s propensity to vote for a party of the extreme right. Finally, we see a significant improvement in the model-fit: the pseudo R2 more than doubles and, more importantly, the BIC is reduced by 1106, meaning that the full model is clearly superior to the socio-demographic one.xviii Given the nature of our explanatory variables, it is also worth noting that multicollinearity is not an issue in our model.xix

 

Starting with the two long-term institutional variables, we can see that the coefficient for the disproportionality of the electoral system is in fact positive, rather than negative as was anticipated.xx That is, the odds of voting for the extreme right actually increase with the disproportionality of the electoral system. At first we considered that this unexpected result might be caused by the inclusion of the French case, where the unique double-ballot system (whose disproportionality scores are extremely high) obviously did not prevent the ascent of the extreme right.xxi We therefore temporarily excluded France from the analysis but found that the coefficient for the disproportionality score hardly changed.

 

The absence of a negative relationship between the disproportionality of the electoral system and the right-wing extremist vote has been reported elsewhere (Carter, 2002), and two potential explanations for it have been put forward: (i) right-wing extremist party voters may simply not be aware of the consequences of electoral systems or (ii) their awareness may be overshadowed by other, more pressing concerns so that the psychological effects of electoral systems have only a weak impact on them. This latter hypothesis has clearly yet to be investigated.

 

As concerns the degree of decentralization and federalism, the coefficient is negative. However, since the coefficient fails the significance test we must accept that our data simply do not provide conclusive evidence as to which of the two hypotheses advanced above holds true in practice.

 

 

Table 3: Complete model

 

 

Independent Variables

b

eb

Male

0.471**

1.602**

(0.042)

(0.068)

Age: -24

0.364**

1.439**

(0.084)

(0.120)

Age: 25-34

0.084

1.087

(0.068)

(0.074)

Age: 35-44

-0.096

0.909

(0.085)

(0.077)

Age: 45-54

-0.200*

0.819*

(0.093)

(0.076)

Age: 55-64

-0.148

0.863

(0.115)

(0.099)

No/Primary Education

0.571**

1.770**

(0.169)

(0.300)

Mid-School Education

0.753**

2.123**

(0.101)

(0.215)

Secondary School Education

0.600**

1.822**

(0.128)

(0.234)

Professionals/Managers

0.007

1.007

(0.267)

(0.269)

Routine Non-Manual

0.082

1.085

(0.207)

(0.225)

Self-employed

0.265

1.304

(0.205)

(0.268)

Manual

0.361

1.435
(0.201) (0.288)

Disproportionality

0.073** 1.076**
(0.017) (0.018)

Index of Decentralisation

-0.116 0.890
(0.132) (0.117)

Ideo. position of major party of mainstream right

0.087 1.091
(0.045) (0.049)

Distance between major parties of mainstream left/right

0.058 1.060
(0.033) (0.035)

Grand Coalition

0.699* 2.011*
(0.356) (0.715)

Asylum Seekers per 1000 inhabitants

0.114 1.121
(0.077) (0.087)

Asylum Seekers: Change

-0.000 1.000
(0.000) (0.000)

Unemployment Rate (%)

-0.222** 0.801**
(0.045) (0.036)

Unemployment Rate: Change

0.006 1.006
(0.005) (0.005)

Constant

-2.439**

(0.148)

Observations

50 276

Pseudo-R2 (Mc-Fadden)

0.07

BIC

-516 399

 

Notes:

Robust standard errors adjusted for clustering on country are shown in parentheses (see note Error: Reference source not found).

* significant at 5%; ** significant at 1%

 

 

Turning to the medium-term party system variables we can see that the position of the major party of mainstream right has a positive and borderline-significant (p = 0.05) effect on the right-wing extremist party vote. A move to the right by the major party of the mainstream right raises the odds of an extreme right vote. This suggests that the second hypothesis advanced above (that a mainstream right party may legitimize the policies of the extreme right by adopting some of their positions) has some validity.

 

The findings also show a positive effect of the distance between the mainstream parties on the right-wing extremist party vote, which is in line with the second hypothesis put forward above. However, since the coefficient does not pass the conventional threshold of significance (p = 0.08), though they are suggestive, our data do not provide conclusive evidence as to which of the competing hypotheses is borne out in practice.

 

The final medium-term party system variable that we included in our model was one that referred to the coalition format of the party systems under investigation. Our findings in Table 3 show that the existence of a grand coalition government before the election in question does indeed have a substantial effect. As we anticipated, the presence of such a governing coalition raises the odds of voting for the extreme right. Depending on the level of the other variables, the probability of an extreme right vote is roughly doubled.

 

As concerns the variables that related to short-term contextual factors, table 3 shows that the effect on the extreme right vote of the number of asylum-seekers is in line with the expectations (it is positive), while the coefficient for the change in the number of asylum seekers is negative. However, both these variables miss the usual threshold for statistical significance by a considerable margin. Therefore, we must assume that their true effect is zero.

 

The effect of unemployment (as a macro variable) on extreme right voting is markedly negative – that is, the odds of voting for the extreme right fall as the rate of unemployment increases. While this clearly does not allow us to draw any conclusions about the extreme right’s appeal to unemployed people (since this would be an instance of ecological fallacy),xxii we can surmise that extreme right parties perform better at the polls in societies where unemployment is low.

 

Although similar results have been reported in other studies (e.g. Knigge, 1998; Coenders and Scheepers 1998, Lubbers et al., 2002), a substantial explanation for this finding is not readily given. One plausible (yet untested) reason for this negative relationship is that people may turn to the more established and experienced mainstream parties in times of economic uncertainty rather than to the parties of the extreme right that lack such experience (Knigge, 1998: 269-270). The coefficient for the change in the unemployment is positive but is not statistically significant, thus again implying that the true impact of this variable on the likelihood of a vote for the extreme right is zero.

 

In the same way that we summarized the findings of our socio-demographic model in Table 2, Tables 4a and 4b summarize the findings of our complete model and show the combined impact of the four strongest system-level predictors on two segments of the population. Table 4a depicts the expected probability of an extreme right vote of a group that is least likely to support parties the extreme right (female voters, aged 45-54, with university education, and from the ‘unclassified’ class category); and Table 4b shows estimates for a small, marginal segment of the general population among which the extreme right is usually quite successful (male manual worker, aged 24 or younger, with no or primary education only).

 

Tables 4a and 4b show the expected probability of an extreme right vote from these two types of voters in situations where:

  1. there is a grand coalition in place in the preceding period of government and when there is not,
  2. the disproportionality of the electoral system is 1 (low) and where it is 5 (high),
  3. the ideological position of the major party of the mainstream right is –5, –1, 1 and 3 (with -5 indicating a rather left-wing position and 3 indicating a more right-wing position), and
  4. the unemployment rate is 2 percent, 4 percent, 6 percent, 8 percent, 10 percent and 12 percent.

 

First, we note that the socio-demographic variables have a considerable and consistent impact even if we control for system-level variables. If we compare equivalent cells from Table 4a and Table 4b, it is obvious that independent of the socio-political context, the probability of an extreme right vote is about five to six times higher for the young male, primary-educated worker than for the mid-aged, unclassified, university educated female voter.

Table 4a: Predicted probabilities (in percent) of an extreme right vote, depending on various system-level variables. Female voters aged 45-54, with university education, and from the ‘unclassified’ class category.

 

 

Female, class unclassified, university education, aged 45-54

Grand Coalition: No

Disproportionality: 1

Disproportionality: 5

Ideo Pos of MRUnempl Rate

-5

-1

1

3

-5

-1

1

3

2

3

5

5

6

4

6

7

8

4

2

3

4

4

3

4

5

6

6

1

2

2

3

2

3

3

4

8

1

1

1

2

1

2

2

2

10

1

1

1

1

1

1

1

2

12

0

1

1

1

0

1

1

1

Grand Coalition: Yes

Disproportionality: 1

Disproportionality: 5

Ideo Pos of MRUnempl Rate

-5

-1

1

3

-5

-1

1

3

2

6

9

10

12

8

12

13

16

4

4

6

7

8

6

8

9

11

6

3

4

5

5

4

5

6

7

8

2

3

3

4

2

3

4

5

10

1

2

2

2

2

2

3

3

12

1

1

1

1

1

1

2

2

 

Notes:

Other system-level variables are held at their respective means, which were calculated giving equal weight to every election.

Typical 95%-confidence intervals: no grand coalition, unemployment 6 percent, disproportionality 1, ideological position of major party of the mainstream right -1: 1.3 – 2.9; grand coalition, unemployment 2 percent, disproportionality 5, ideological position of major party of the mainstream right 1: 6.9 – 24.4.

 

 

This said, the impact of the system-level variables is considerable, too. Depending on the variable constellation, the presence of a grand coalition government before the election almost doubles the support for the extreme right (to see this, compare equivalent cells in the upper and lower parts of either Table 4a or 4b). The position of the major party of the mainstream right has almost the same impact: if it is closer to the empirical right end of our scale, the probability of a vote for the extreme right is about 1.5 to 2 times higher than in situations where this party is further to the left of our scale. To see this effect, we can look at each row and compare the first and the fourth, and the fifth and the eighth cell respectively.

 

 

Table 4b: Predicted probabilities (in percent) of an extreme right vote, depending on various system-level variables. Male manual workers, aged 24 or younger, with no or primary education only.

 

 

Male, manual, no/primary education, aged 24 or younger

Grand Coalition: No

Disproportionality: 1

Disproportionality: 5

Ideo Pos of MRUnempl Rate

-5

-1

1

3

-5

-1

1

3

2

20

26

29

33

25

32

36

40

4

14

18

21

24

17

23

26

30

6

9

12

14

17

12

16

19

21

8

6

8

10

11

8

11

13

15

10

4

6

7

8

5

7

9

10

12

3

4

4

5

3

5

6

7

Grand Coalition: Yes

Disproportionality: 1

Disproportionality: 5

Ideo Pos of MRUnempl Rate

-5

-1

1

3

-5

-1

1

3

2

33

41

45

50

40

48

53

57

4

24

31

35

39

30

37

42

46

6

17

22

25

29

21

28

31

35

8

11

16

18

21

15

20

23

26

10

8

11

12

14

10

14

16

18

12

5

7

8

10

7

9

11

13

 

Notes:

Other system-level variables are held at their respective means, which were calculated giving equal weight to every election.

Typical 95%-confidence intervals: no grand coalition, unemployment 6 percent, disproportionality 1, ideological position of major party of the mainstream right -1: 10.8 – 14.5; grand coalition, unemployment 2 percent, disproportionality 5, ideological position of major party of the mainstream right 1: 47.8 – 57.7.

 

 

By contrast, the effect of disproportionality is rather moderate: the probability of a vote for the extreme right is 1.1 to 1.5 times higher in a situation in which there is high disproportionality than in a situation where disproportionality is low. We can see this if we compare the left and the right halves of Tables 4a and 4b.

 

Lastly, our model shows that unemployment has a massive impact on the probability of a vote for the extreme right. A two percentage point increase in the unemployment rate reduces the probability of a vote for the extreme right by between one third and one fifth (depending on the other variables). To see this, we can compare any cell in Table 4a or 4b with the cell directly above or beneath it.

 

The combined impact of these four system-level variables alone is large – something which becomes obvious if we compare a situation where, according to our findings the extreme right should be least successful (i.e. high unemployment, no grand coalition, low disproportionality, major mainstream right far to the left) with a situation where the extreme right is expected to be most successful (i.e. reversed conditions). In situations of the first type, our prototypical female voter has an expected probability of voting for the extreme right of (almost) 0 percent. By contrast, in situations of the second type, this same voter has a predicted probability of voting for the extreme right of 16 percent. In other words, when we compare the two situations, the expected probability of an extreme right vote from our female voter varies by a factor of about 40.

 

If we look at the expected probability of an extreme right vote from our male voters in the two different situations, we expect a support of 3 percent in a situation where the extreme right is expected to be least successful, and a support of 57 percent in a situation where the extreme right is expected to be most successful. The expected probability of an extreme right vote from our male thus varies by a factor of roughly 22.

 

Clearly, these probabilities are open to interpretation and should not be seen as set in stone as our model does not fit the data perfectly, is based on only 24 elections, and might not contain all the relevant system-level predictors even though our range of variables is considerably broader than in previous analyses of the extreme right vote. Furthermore, our scenarios are somewhat counterfactual in that, in the past, all the conditions that according to our model favour the extreme right have never been present simultaneously in one country – and neither have all the conditions that seem to hinder the success of the parties of the extreme right. Therefore, in reality, there would probably be a limit to the potential of the parties of the extreme right, whereas our model assumes that the effects of the system-level factors are additive in the logits. This said, however, even if we take the probabilities estimated by our model as guidelines rather than exact prognoses of an extreme right vote, they nonetheless provide clear testimony to the importance of system-level factors in explaining the probability of an extreme right vote, and hence in accounting for uneven electoral success of the extreme right across the countries of Western Europe.

 

 

Conclusion

 

In the course of our analysis, we have shown that a voter’s socio-demographic attributes go a long way towards explaining his or her propensity to vote for a party of the extreme right. Our results – which confirmed many of the conclusions reached in the existing country studies – indicate that being male, being young (under 25), and being a manual worker significantly raised the probability of voting for the extreme right in all the elections under study, whereas being female, being in the middle age categories and being a professional markedly decreased the probability of voting for a party of the extreme right. The only slightly unanticipated result was the finding that voters with mid-school levels of education (rather than those with lower levels of education) who had the highest propensity to vote for the extreme right.

 

However, although our results provide a good basis for predicting the likelihood of an extreme right vote, socio-demographic characteristics do not go very far in explaining why the parties of the extreme right have encountered greater levels of electoral success in some instances but have experienced relative failure in others. Therefore, we estimated an augmented model that allowed us to assess the degree to which political opportunity structures account for the variation in the extreme right’s vote after individual-level socio-demographic characteristics had been controlled for.

 

The impact of system-level variables is considerable. In particular, our results show that the level of unemployment, the position of the major party of the mainstream right, the disproportionality of the electoral system, and the presence of a grand coalition government are particularly important in explaining the uneven success of the right-wing extremist parties across Western Europe. The effects of most of these variables were as we anticipated: we found that the more to the right the mainstream right party, the greater the likelihood of an extreme right vote being cast, suggesting that a right-wing mainstream party may have a legitimizing effect on the policies of the extreme right. Our findings also showed that the presence of a grand coalition government prior to elections raises the odds of an extreme right being cast, most probably because levels of voter dissatisfaction are higher during periods of grand coalitions than during periods of alternating government.

 

By contrast, some of our other results defy common wisdom: we found that the coefficient for the disproportionality of the electoral system was in fact positive, suggesting that right-wing extremist voters are not responding to the psychological effects of electoral systems in the way one might expect. In addition our results showed that the effect of unemployment (as a macro variable) was markedly negative, perhaps because voters turn (back) to the more experienced mainstream parties in times of high unemployment.

 

Therefore, we believe that, above and beyond their academic worth, our findings have implications for the real world. In particular, they suggest that the ring-wing extremist vote will not be curbed by simply looking after economic conditions. They also indicate that tampering with electoral systems (to render them less proportional) might not lead to lower extreme right party scores. Furthermore, our results imply that, in the West European case at least, a move to the right by a party of the mainstream right is more likely to legitimize the extreme right than quell the demand for the latter’s policies. These findings thus go some distance towards challenging the conventional wisdom as to how the advance of the parties of the extreme right may be halted.

 

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iNotes

 

 Their data contain a very low number of level two units (countries). According to much of the literature, the number of level-two units should be at least 30 (Snijders and Bosker, 2000: 140; Kreft and de Leuw, 1998: 124-5; Hox, 2002: 173-9), and if one is interested in the variance components (as Lubbers et al. are), then this number should be even higher (Hox, 2002: 175).

ii We do not incorporate the positions of the parties of the extreme right in our model because (i) we are above all interested in the space available to the right-wing extremist parties (ii) including both space and positions would lead to problems of multicollinearity and (iii) because some of these parties are not included in the CMP data.

iii Although the position of the major party of the mainstream right and the ideological convergence between the tow major parties are conceptually related, the empirical correlation between both measures is negligible (r=-0.18).

v Since all our Belgian extreme right voters voted for the Vlaams Blok, in this case our party variables actually relate to the Flemish party system.

vi Despite our best efforts, we were forced to exclude Sweden, the Netherlands, and Switzerland from our analysis, as we were unable to access national election studies in these countries.

vii There is substantial agreement within the literature that after the Fiuggi Congress in 1995, the AN gradually became a part of the mainstream right (Newell, 2000). In light of this, the post-1995 AN is not included in our analysis.

viii Indeed, in these countries not a single respondent out of several thousand reported having voted for an extreme right party (see Lubbers et al., 2002: 357).

ix Three of the elections under study (Austria 1999, Belgium 1999 and Norway 2001) took place after the CMP data were gathered. For these, we made use of the positions of the parties at the most recent election for which CMP data do exist.

x For aggregate level data see LABORSTA (http://laborsta.ilo.org) and Statistics Norway (http://www.ssb.no/english/subjects/06/01/aku_en/).

xi Data for 2001 were obtained from the UNHCR (http://www.unhcr.ch), Data for all other years are from OECD-SOPEMI.

xii We chose to use this figure because (i) when asked about ‘foreigners’, the majority of citizens in the countries under study think of people from outside Western Europe (Fuchs et al., 1993) and (ii) the alleged ‘flood’ of refugees and asylum-seekers from outside Western Europe became the main target of the extreme right’s appeals in the countries under study.

xiii Variations in the ideology of the extreme right could be seen as the exception to this statement. However, since information on their position is not available for all parties and elections (see note Error: Reference source not found) and since we are primarily concerned with exogenous conditions for their success, we treat all variation in the strength of the effects as random error.

xiv Throughout this paper we report ‘robust’ standard errors, which correct for heteroscedasticity and adjust for correlated disturbances within countries, thereby yielding very conservative t-statistics and confidence intervals.

xv The coefficients for those in the three middle age categories are jointly different from the reference group although two of them fail individual significance tests.

xvi The last three coefficients are jointly significant although they fail the individual tests.

xvii The one notable exception to this is the coefficient for no/primary education, which is now closer to the coefficient for mid-school.

xviii The BIC reflects the trade-off between model fit and loss of degrees of freedom. A difference of10 ore more is regarded as lending very strong support to the model with the smaller BIC.

xix See Arzheimer and Carter (2003: 31) for further details.

xx This variable reflects the disproportionality score of the previous election.

xxi See Arzheimer and Carter for a more elaborate discussion of this point.

xxii Unfortunately, individual unemployment it is not consistently recorded in the surveys.

Die Wähler der Extremen Rechten 1980-2002

 

Spätestens seit den frühen 1980er Jahren haben sich die Parteien der Extremen Rechten – manchmal auch als Radikale Rechte, Neue Rechte oder Populistische Rechte bezeichnet – als neue Parteienfamilie in Westeuropa etabliert. Fast jeder der alten EU-Mitgliedsstaaten sowie Norwegen und die Schweiz mußte sich in diesem Zeitraum mit einer oder mehreren dieser Parteien auseinandersetzen, deren Verhältnis zur liberalen Demokratie häufig als problematisch erscheint.

Die Unterstützung für diese Parteien schwankt allerdings in erheblichem Umfang über die Zeit und zwischen den hier untersuchten Ländern. Es liegt nahe, dies auf politische Faktoren zurückzuführen, die sich der Kontrolle der Rechtsparteien weitgehend entziehen. In dieser Studie werden das Wählerprofil der Extremen Rechten sowie der Einfluß von Kontextfaktoren auf deren Wahlerfolge untersucht.

Contextual Factors and the Extreme Right Vote in Western Europe, 1980-2002

 

After the Second World War, the extreme right (ER) in Western Europe was associated with the atrocities of the Nazis and their puppet regimes (Rydgren 2005) and was therefore politically isolated and insignificant in most countries of the region. But from the early 1980s on, an unexpected third wave of right-wing extremist party activity swept over the continent. All of a sudden, parties that were dubbed as “extreme”, “radical”, “populist” or “new” right proved highly successful at the polls in countries such as Austria, Belgium, Denmark, France, Italy, Norway, Sweden, and Switzerland.

Problems of terminology and idiosyncratic features notwithstanding, a consensus1 emerged that these parties should be grouped into a single party family. While this group of extreme right parties (ERP) is arguably more heterogeneous than other party families (Mudde1996), its members are reasonably distinct from the mainstream or established right and share a number of ideological features, in particular their concern about immigration, which swiftly became the single most important issue for these parties (van der Brug and Fennema2003).

By the 1990s, scholars of electoral behavior had also identified a set of core features of the ERP electorates. While the most successful of these parties have managed to attract some votes from virtually all social groups, the bulk of the extreme right’s support comes from non-traditional segments of the working and lower-middle classes who are worried about the presence of Non-West European immigrants in their respective societies. There is generally a much greater propensity to vote for the ER amongst men, voters who are either young or rather old, those with a low level of formal education and amongst the manual workers, the petty bourgeoisie, and those in routine non-manual employment (see the review in  Arzheimer and Carter2006, 421-422). This sharp social profile is matched by an equally clear attitudinal profile: as a number of studies have demonstrated, the voters of the West European ER are to a large degree motivated by xenophobic feelings and beliefs (see e.g. van der Brug and Fennema2003).

A whole host of national and a smaller number of comparative studies have replicated these findings time and again. However, surprisingly little attention has been paid to the more intriguing twin question of why the ER’s support is so unstable within many countries over time, and why these parties are so weak in many West European countries. Only a handful of contributions have looked into this question at all, and each of the existing studies has its shortcomings. Moreover, the findings from different studies often contradict each other. The aim of this article is therefore to employ fresh data and a more adequate modeling strategy in order to provide a more comprehensive and satisfactory answer to the question of why the support for the ER in Western Europe varies so much over time and across countries.

The remainder of this article will proceed as follows. After a brief introduction to the main theories of ER voting, the existing longitudinal and comparative studies on the ER vote in Western Europe will be reviewed. Following that, a multi-level model of the ER vote in Western Europe will be presented. The article ends with a discussion of the main findings and their implications for future research.

Theoretical Accounts for Extreme Right Support in Western Europe

Starting with then contemporary attempts to explain the rise of the Nazi party and the Italian Fascists, social scientist have developed a multitude of theoretical accounts of support for the extreme right. The complexity of these accounts notwithstanding, they can be grouped in four broad strands (Winkler1996).

A first group of authors focuses on largely stable and very general attributes of the ER’s supporters, namely personality traits and value orientations, that make them more receptive for the ER’s appeals than their compatriots. The most prominent example of this line of research is arguably Adorno et al. (1950).

A second strand of the literature is chiefly concerned with social disintegration, which is characterized by a (perceived) break-down of social norms (“anomia”) and intense feelings of anxiety, anger and isolation brought about by social change. Allegedly, this mental state inspires a longing for strong leadership and rigid ideologies that are provided by the ER. A classic example of this approach is Parsons (1942).

According to a third class of accounts which draw heavily on theories from the field of social psychology, group conflicts are the root cause of the ER’s successes. This strand of research is, however, rather heterogeneous. At one end of the spectrum, there are classical theories of scapegoating (e.g. Dollard et al.1939). They argue that (ethnic) minorities provide convenient targets for the aggression of those members of the majority who are frustrated by their lack of status and other resources because these minorities tend to be both different from one’s own reference group and powerless. Otherwise, the choice of the victimized group is largely random and purely driven by emotions.

At the other end of the range, theories of Realistic Group Conflict beginning with Sherif and Sherif (1953) emphasize that ethnic conflicts can be driven by a bounded yet instrumental rationality. If xenophobia is the result of a conflict between immigrants and lower class natives over scarce resources (low-paid jobs, welfare benefits), discrimination against immigrants, proliferation of racist stereotypes and support for the ER can be interpreted not as an emotive reaction but rather as part of an instrumental strategy. This idea is especially prominent in more recent accounts (e.g. Esses, Jackson and Armstrong1998).

Finally, theories of ethnic competition (Bélanger and Pinard1991), “status politics” (Lipset and Bendix1951) , “subtle”, “modern” or “symbolic” racism (Kinder and Sears1981), and social identity (Tajfel et al.1971) all cover a middle ground between these two poles. While the various labels obviously highlight different aspects of group conflicts, recent research (Pettigrew2002) usefully suggests that most if not all of these approaches could be subsumed under the concept of relative deprivation: members of one social group feel that in comparison with another social group, they are not getting what they feel they are entitled to, even if they know that they get more than the other group.

While all three approaches have a lineage that spans more than five decades, most recent comparative research explicitly or implicitly combines theories of group conflict with elements of a fourth perspective that complements and expands on the three major approaches. In Winkler’s (1996) original survey of the literature, this emerging perspective was presented under the label of a “political culture” that constrains the effects posited by the other approaches. However since the mid-1990s, interest in a whole host of other, more tangible contextual factors has grown tremendously, and it is now widely believed that the interplay between group conflicts and system-level variables can help explain the striking differences in support for the ER over time and across countries.2

Building on previous work by Kriesi et al. (1992) and Tarrow (1996), Arzheimer and Carter (2006) have argued that these contextual factors should be subsumed under the concept of a “political opportunity structure” . Such a structure consists of short-, medium- and long-term contextual variables, which capture the “openness or accessibility of a political system for would-be political entrepreneurs” (Arzheimer and Carter2006, 422) and affect the chances (and thereby presumably the motivation) of politicians to create and maintain an electorally viable ERP.

However, while the concept of opportunity structures is certainly useful, it is also notoriously vague. As the review in the next section will demonstrate, there is no consensus (yet) on what variables are part of an opportunity structure. On the other hand and somewhat paradoxically, the notion of “opportunity” has implications that might be to restrictive: many context factors like unemployment or immigration will not only provide the political elite with an incentive to mobilize as entailed by the concept but will also have a direct and possibly more important impact on voters’ preferences. Given that comparative data on the perceptions and strategic decisions of (would-be) members of the political elite are unavailable, it is empirically impossible to separate the two potential effects of a contextual variable.

More generally, social psychological theories of group conflict were developed and tested in the context of small group research, where psychological and sometimes even physiological processes can be closely monitored, often in an experimental or quasi-experimental setting. Datasets that are available for longitudinal and comparative analyses, on the other hand, are restricted to a handful of attitudinal measures and a set of simple socio-demographic variables that were consistently replicated over the years. This problem is mitigated, however, by the fact that all theoretical accounts of ER support tend to identity a similar set of socio-demographic groups that should be most susceptible to the appeals of the ERPs. Moreover, both national and cross-sectional comparative studies of ER support have confirmed strong and consistent links between these socio-demographic indicators of group membership and more nuanced attitudinal measures.

Therefore, although data limitations make is impossible to unpack the details of the underlying psychological process, it is clear that the impact of micro- and macro-level variables on support for the extreme right in Western Europe should be modeled jointly. Only such a multi-level model provides one with unbiased estimates of the contextual effects, because the differences in the socio-demographic and attitudinal composition of the European electorates are controlled for. A multi-level model therefore represents a significant improvement over the existing empirical accounts for the ER’s support, which will be reviewed in the next section.

Previous Findings on Contextual Determinants of the Extreme Right’s Electoral Support

Jackman and Volpert (1996) conducted the first large scale3 quantitative comparative analysis of the ER’s electoral support by estimating a Tobit model of the ERPs’ vote share. Their main findings were that(1) the ER benefits from high unemployment,(2) higher electoral thresholds reduce the support for the ER, and(3) multi-partyism in combination with a proportional electoral system is associated with higher levels of ER voting.

Some technical issues notwithstanding (see Golder2003a), the analysis by Jackman and Volpert was ground-breaking both in terms of its spatial and its temporal coverage, and yet, there are some obvious substantive problems with it. First, Jackman’s and Volpert’s selection of cases is problematic in at least one instance: They include the Alianza Popular/Partido Popular, which became the major party of the established right in Spain in the 1980s and is not normally considered an ERP (Ignazi2003, 190-191). On a related account, their time-frame is problematic since there is now a wide agreement that the “third wave” did not come to itself before the early 1980s, when immigration became the core issue of the ER and many ERP tried adopted new strategies and communication frames that had proved successful in France (Rydgren2005). This leads to an obvious problem with the Scandinavian Progress Parties that started out as anti-tax parties in the 1970s and only moved into the ER camp during the early 1980s (Svåsand1998).

Second, Jackman and Volpert (1996) analyze the impact of a (somewhat limited number of) polity-level variables on aggregate support for the ER but completely ignore the micro-level, which is at the center of all theoretical explanations of ER voting. Finally, by modeling electoral returns, Jackman and Volpert restrict their analysis to a handful of (very important) snapshots in the political histories of the 16 countries. But while election results are decisive for the creation, composition and survival of governments, the ongoing level of support for the ER can have a tremendous impact on proposed and actually implemented policy via the strategic calculations of the established parties, even if the ER is not (yet) represented in parliament (Minkenberg2001).

For these reasons, Knigge (1998) models aggregate support for ERPs as measured by the bi-annual Eurobarometer surveys in Belgium, France, the Netherlands, West Germany, Denmark, and Italy between 1984 and 1993 in a Time-Series Cross-Sectional design and concludes that immigration and political dissatisfaction correlate with higher levels of support for the ER. Conversely, the effect of unemployment is negative.

While Knigge’s contribution is an improvement over the analysis by Jackman and Volpert because she analyses time-series with more and uniformly spaced data-points, it clearly falls short in terms of country-coverage. Moreover, like Jackman and Volpert, Knigge confines herself to the macro-level while a comprehensive model of extreme right support should clearly include both micro- and macro-level factors.

Precisely this is the aim of the useful study by Lubbers, Gijsberts and Scheepers (2002), who merge surveys from 16 West European countries with a host of aggregate variables. From a series of multi-level models they conclude that after controlling for individual anti-immigrant attitudes and political dissatisfaction, the number of non-Western residents as well as characteristics of the ERP themselves have a substantial impact on the likelihood of an extreme right vote, whereas the unemployment rate has no significant effect.

However, their contribution is problematic, too, in a number of ways. First, the merging of data from six national election studies with data sets from three different supra-national projects obviously raises problems of validity and reliability. Second, the number of N = 17 level-two units is too low for multi-level modeling by any conventional standard, especially given the authors’ interest in estimating variance components (Hox2002, 173-179). Finally, they focus on a rather brief time period, thereby excluding the 1980s (when the ER became a relevant political actor for the first time in decades) and much of the 1990s. Moreover, unlike Jackman and Volpert and Knigge, Lubbers, Gijsberts and Scheepers discard any cross-temporal variation in the ER’s support by pooling surveys from different years.

This is certainly not a problem in the analysis presented by Golder (2003b), which proceeds along similar lines like that of Jackman and Volpert but covers 19 West European countries including many “failed cases” like Iceland, Ireland, or Malta, and 165 elections between 1970 and 2000. From his findings, Golder concludes that(1) the ER benefits from both high levels of unemployment and high levels of immigration, and that(2) there is an additional positive interaction between unemployment and immigration..

Although Golder’s results are suggestive, like the other aggregate analyses they do not allow one to draw conclusions about micro-level processes, e.g. about the propensity and the motivation of the unemployed to vote for the ER. Technical sophistication notwithstanding, most of the problems discussed in regard with Jackman and Volpert and Knigge therefore apply to Golder’s study as well.

While all studies discussed so far have addressed unemployment as one potential determinant of ER support, Swank and Betz (2003) were the first who empirically analyzed the the mediating effects of welfare state institutions on the ER vote. In yet another macro model, they regress the electoral returns of the ER in 83 elections that were held between 1981 and 1998 in 16 West European countries on trade openness, capital mobility and foreign immigration as well as on the level of social protection and a number of other contextual variables. From their findings, they conclude that the number of asylum seekers is positively related to ERP success, whereas a high level of welfare state protection reduces the appeal of the ER.

However, although the impact of unemployment is at the center of the debate, Swank and Betz (2003, 228) use a fairly general index of welfare state benefits. While their approach is innovative, this variable is clearly not ideal for their purpose. Moreover, all concerns regarding the aggregate analyses by Jackman and Volpert, Knigge, and Golder obviously apply here, too.

Finally, in the most recent contribution to the field, Arzheimer and Carter (2006) have tried to overcome some of the limitations of the existing research by merging data from 24 national election surveys (conducted in seven countries between 1984 and 2001) with a host of aggregate-level information such as party-positions, proportionality of the electoral system, unemployment, and immigration figures. Like Knigge, they find a negative effect of aggregate unemployment. Moreover, they conclude that established right-wing parties that take a very tough stand on immigration may actually legitimize the policies of the ER, and that grand coalition government prior to elections raises the odds of an ER vote being cast.

However, Arzheimer’s and Carter’s paper is not without problems, too. First, while their mode of analysis is less demanding on the data than multi-level modeling, the parsimony of their model comes at a price as they have to assume that no unit (=country) effects remain after controlling for the impact of the aggregate variables. If this assumption does not hold, bias will result. Second, unlike Knigge and Lubbers, Gijsberts and Scheepers, they do not measure support for the ER between elections. Third, since they rely on national election studies that vary wildly in terms of attitudes questions asked, their range of individual-level variables is restricted to “objective” features like age, class, and education. Fourth, the number of level-2 units (elections) is rather low in relation to the large number of contextual variables in which they are interested.

To summarize, the existing research has demonstrated that contextual factors (and most prominently immigration and unemployment) have a systematic effect on support for the ER in Western Europe. However, while models of ethnic competition (that are consistent with micro-level theories of support for the ER) strongly suggest that immigration, unemployment, and their interaction should all have positive effects, it is unclear whether and under what conditions this is true. Moreover, it is by no means obvious that unemployment and immigration are truly more important than other contextual factors.

Support for the Extreme Right in Western Europe, 1980-2002

Model

While a lot of progress has been made since the paper by Jackman and Volpert was published, the previous section has shown that none of the existing studies on the contextual determinants of the extreme right’s vote is entirely satisfactory. The analysis presented here tries to overcome these limitations by(1) combining a relatively large number of relevant system-level variables with individual socio-demographic and attitudinal data that are measured in a comparable fashion,(2) covering the whole time-span between 1980 and 2002, and(3) not excluding contexts where the ER is very weak.

At the micro-level, the model includes information on the respondents’ gender, age, level of education, and social class. The gender gap in support for the ER is well known, even if its causes are controversial (Gidengil et al.2005). The other socio-demographic indicators reflect the theoretical and empirical links between group membership, attitudes and the likelihood of a vote for the ER. For instance, it is well known that voters with high levels of educational attainment are more likely to embrace liberal values (Weakliem2002) and have little reason to feel threatened by low-skilled immigrants. Younger voters, members of Europe’s declining “petty bourgeoisie”, manual workers, the unemployed and maybe pensioners, on the other hand, should be highly susceptible to the appeal of the ER because they compete with immigrants for scarce resources.

Three measures model the impact of ideologies and more specific political preferences. First, while longitudinal data on immigration attitudes are largely unavailable, in most countries the ER has taken a negative stance on European Integration and has tried to link this theme to its core issues of immigration, national sovereignty, and law and order. Therefore, the model contains a control for Euroscepticism.

Second, the notion of a “protest vote” features prominently in some of the earlier accounts of the “third wave”. It is, however, unclear what a protest vote should entail. On the one hand, some authors suggest that “protest” is something irrational and emotional that is unconnected to values and ideologies and primarily “a vote against things” (Mayer and Perrineau1992, 134). But on the other hand, it is obvious that much of this “protest” is not un-ideological at all but clearly directed “against the policy or the absence of policy in this respect [migrants and law and order]” (Swyngedouw2001, 218-219).

To account for these “protest motives”, the model contains both an indicator for general political dissatisfaction as well as a control for political ideology (the familiar left-right self-placement scale). This makes it possible to separate the alleged “pure protest” from ideology- and policy-based considerations. Moreover, controlling for ideology accounts for the fact that the political left can benefit from Euroscepticism, too.

Most national and comparative studies of the ER vote have demonstrated rather stable and uniform effects for these individual-level variables. The crucial question here is whether these regularities do still hold once contextual variables are included and the spatial and temporal coverage is extended beyond that of previous analyses.

At the macro-level, the model aims at bringing together the most relevant variables from the contributions discussed above without taking a shotgun approach that would render both estimation and interpretation infeasible. Given their prominence in the literature and the fact that ethnic competition theories provide a clear rationale for the interpretation of their effects, the inclusion of unemployment and immigration plus an interaction between both figures is a matter of course.

The “protective” effect of welfare state benefits found by Swank and Betz (2003) deserves closer inspection,too, not least because it has clear implications for public policy. Moreover, the findings by Swank and Betz contravene one particularly influential early account of the ER’s support in Western Europe: Kitschelt’s (1995) hypothesis that a combination of authoritarian and market-liberal stands would guarantee electoral success for the ER. However, in line with the argument about ethnic competition in the labor market, instead of the general benefit data a more specific measure of benefits for the unemployed will be used.

Following Arzheimer’s and Carter’s approach, two institutional features that most clearly embody the concept of a (durable) opportunity structure, namely political decentralization and the degree of disproportionality of the electoral system are included, too. In the case of decentralization, Arzheimer and Carter present arguments both for a positive and a negative relationship with the ER vote. On the one hand, subnational elections can work as a “safety valve” for dissatisfied citizens that would ceteris paribus reduce support for the ER in national elections. On the other hand, these second-order elections provide the ER with opportunities for acquiring political experience, access to the media, and credibility. While neither of these two effects is borne out in their original analysis, disproportionality is of particular interest because the existing research seems to disproves the common wisdom that less proportional systems help to “keep the rascals out”.

Finally, Lubbers, Gijsberts and Scheepers as well as Arzheimer and Carter have argued that a comprehensive model of ER voting should also reflect the impact of genuinely political short-term factors such as the political agenda, the general tenor of the political debate on the issues of the ER, and the ideological positions of the political parties in a given country at a given time. While both groups of authors use a somewhat idiosyncratic terminology, their two competing hypotheses can easily be re-expressed within the framework of well-established theories of political behavior.

On the one hand, classical theories of spatial competition (cf. Enelow and Hinich1984) which treat the distribution of preferences in the electorate as exogenous and fixed in the medium run suggest that the rise of ER in the 1980s can be explained by the mainstream parties’ persistent reluctance to cater to the existing demand for strict immigration and asylum policies. Consequentially, support for the ER must decline if the established parties (with their track record of past performance in government and their much broader appeal) take a tougher stand on immigration and multi-culturalism, thereby “stealing” the ER’s issues. As Bale (2003, 76) observes, this “conspiracy of silence” theory of ER electoral successes is rather popular with political pundits in many West European countries.

On the other hand, a more subtle claim is often made in the literature on the extreme right: if a mainstream party takes a radical position on the extreme right’s issues, the public can interpret this as a signal that these policies are relevant, and that the contents and style of extreme right politics are no longer taboo (see e.g. Thränhardt 1995). As a consequence, at least some of the voters who support the ER’s policies but shy away from voting for a stigmatized party will now, in Jean-Marie Le Pen’s words, “prefer the original to the copy”. Moreover, other voters who were previously not aware of these issues may now be induced to evaluate the parties with respect to this policy dimension.

While authors like Thränhardt and Bale interpret this chiefly as a Machiavellian gambit by the established right (who have less to lose and more to win than the established left if immigration moves up the political agenda), an increase in the importance of immigration, asylum and race will without doubt benefit the extreme right, too. Although the connection is rarely made in the literature, these effects can easily be interpreted in terms of agenda setting and priming

To gauge the potential effects of party competition, Lubbers, Gijsberts and Scheepers (2002) rely on an expert survey, from which the derive two measures that capture the “immigration restriction climate” and the available “space for the ER”. Arzheimer and Carter (2006) draw on the Comparative Manifesto Dataset. Using party statements on internationalism, multi-culturalism, national lifestyle, and law and order, they construct two variables, namely the ideological position of the major party of the ER and the ideological distance between the two major mainstream parties. The latter approach seems preferable, because unlike the expert survey, the manifesto data are inherently dynamic, based on a well-defined and reliable coding procedure, easily available for replication, and cover the whole period under study.

However, to further improve on Arzheimer and Carter and to link the empirical analysis more closely to the underlying theories, the construction of both variables was slightly modified. First, considering only the ideological position of the major moderate right party seems overly restrictive. Often, the margin of what is politically acceptable will in fact be defined by the position of a smaller party of the right (or left, see Thränhardt 1995, 328 on anti-immigrant measures taken by Communist mayors in France).4 Consequentially, the most radical position on the ER’s issues taken by an electorally relevant party that is not part of the ER (cf note 11) is used as an indicator for electoral competition. This approach has the additional benefit of avoiding somewhat arbitrary decisions about what constitutes the “major” party amongst a whole group of more or less equal-sized political groupings.

Second, Arzheimer’s and Carter’s indicator of convergence between the two major parties was replaced by two separate measures for the variance and the salience of statements by all established parties on the issues of the ER. The salience measure ignores the direction of these statements and focuses solely on the space devoted to these issues. More salience is equivalent to a more prominent position of these issues on the agenda, which presumably benefits the ER. The variance measure, on the other hand, reflects Zaller’s (1992, chapter 6) more subtle proposition that the public will often follow the views of the elites if there is consensus amongst them, whereas visible disagreement amongst the elites conducive to polarization.5

To summarize, if standard spatial theories of voting apply, support for the ER will ceteris paribus be lower where the established parties position themselves further to the right. If, however, theories of agenda setting and priming prevail, the extreme right should benefit if their issues(1) feature prominently in elite discourses directed at voters and(2) if there is little consensus about what should be done. On the other hand, if elites downplay these issues and if there is little conflict amongst elites, this should reduce support for the ER. Consequentially, the “conspiracy of silence” could be a viable political strategy.

Finally, two caveats are in order. First, the phrase “no longer taboo” suggest that the timing of manifesto statements can have a crucial effect: once a taboo is broken, it could be difficult if not impossible to restore it. In principle, one could classify countries according to(1) whether the established parties have ever adopted the issues of the ER and(2) if so, whether they have returned to their original position, and introduce this classification as an additional variable. But since the number of West European countries is low, in reality it is infeasible to model this effect of timing.

Second, the model does not contain any measures for another important class of short-term factors, namely the content of the mainstream-media. However, while the media will most probably haven some effect even when party positions are controlled for (see Boomgaarden and Vliegenthart 2007 for a single-country study that tests this proposition), relying solely on party manifestos can actually be an advantage since it could be argued that political messages sent by other parties could to a degree reflect anticipations about future and reactions to previous successes of the ER, which would in turn lead to endogeneity bias. While this argument may apply to the statements that parties and politicians issue on a daily basis, endogeneity is less likely to be a problem with party manifestos, manifestos are the outcome of a lengthy deliberation process within the respective party. Moreover, unlike individual statements, the commit they represent a public policy commitment. Therefore, manifestos are probably the best and most reliable measure for a party’s position and political message.

Data

The analysis covers the member states of the European Union (EU) as it existed before the Eastern enlargement rounds plus Norway.6 Individual level data come from the European Commission’s bi-annual series of Eurobarometer surveys.7 The number of missing values in the Eurobarometer for the variables under study is rather low, yet listwise deletion of cases with missing information reduces the sample size by about one third and can lead to overly optimistic standard errors and biased estimates. As a safeguard, the Multivariate Imputation by Chained Equations procedure devised by van Buuren and Oudshoorn (1999) was used to create eleven imputed data sets. All analyses were carried out both on the original and the completed data sets, but the results are almost identical.

Contextual information was drawn from official election results, OECD databases and printed reports (OECD1992, 1999, 2001200220032004), the datasets produced by the Comparative Manifesto Project (Klingemann et al.2006), the UNHCR statistical yearbook (UNHCR2002), and Lijphart’s (1999) seminal study of institutional arrangements in Western democracies.

The analysis spans the years from 1980 to 2002. During these 23 years, 1,065 individual Eurobarometer surveys were conducted in 18 countries.8 Each survey where at least one respondent voiced an intention to vote for an ERP and where all the required individual-level information was available was retained, yielding a total of about 175.000 respondents nested in 267 individual surveys. Surveys without any supporters of the ER were excluded, which effectively removed the United Kingdom and the Republic of Ireland from the analysis.9

While Golder (2003b) is right that excluding these “failed cases” is likely to lead to biased estimates in studies where aggregated electoral support is modeled, the case is less straightforward in a setup where individual voting intentions are analyzed. In countries where the ER is very weak, strong effects of social desirability are likely to bias the measurement of ER support. Moreover, supporters of the ER are often prevented from voting for their preferred party because the ERPs will not field candidates in most constituencies.10 Finally, due to financial constraints of the pollsters, the supporters of tiny ERPs are often coded as voting for “other” parties. As a result, support for the ER will be underestimated in contexts where those parties are already very weak, which will lead to a different kind of bias. While there is no perfect solution for this dilemma, restricting the analysis to contexts where it is at all possible to trace support for the ER by means of mass opinion surveys is a reasonable compromise.

The remaining surveys provide an exceptionally good coverage of the “third wave”, including the early successes and failures in the 1980s. The only major gaps are Norway in the 1980s and late 1990s (when the country was not yet/not any longer covered by the Eurobarometer) and Austria between 1986 (when Haider became chairman of the FP”O) and 1994 (when Eurobarometer polling started).

Method

The dependent variable is vote intention for an ERP,11 calling for logit or probit multi-level analysis because the observations in the dataset are obviously nested. However, the way in which this nesting should be modeled is less obvious. Observations could be conceived of as (1) persons nested in countries nested in time, (2)  persons nested in time nested in countries, or (3) persons, cross-classified by time and countries. Of these, (2) is the most appropriate variant for a number of reasons. First, a cross-classification would be structurally incomplete, because a number of country-years are not covered by the Eurobarometer or were excluded because there were no ER voters. Second, persistent effects unit (country) effects are quite strong, whereas there is no indication of any effects of time that would be uniform across countries. Finally, while time-points are random in the sense that they can be conceived of as a sample from a large universe of days/weeks on which a survey could have been conducted, countries are not sampled from a population but are essentially “fixed” (Berk, Western and Weiss1995).

For these reasons, countries are represented by a series of dummy variables, which are common to all surveys from a given country.12 This modeling strategy effectively reduces the number of levels to two (see Duch and Stevenson 2005 for an application), namely the individual and the particular context of the survey wave in which she was interviewed. Thus, the model can be written as

Please see the PDF-Version

where i is an index at the person-level and j is an index at the context level. Hence, yij is the individual vote intention for an ERP, which is assumed to be binomially distributed (??). The logit of the probability to vote for the extreme right (??) depends on a linear combination of k individual variables (x1ij⋅⋅⋅xkij), l contextual variables (z1j⋅⋅⋅zlj), 14 fixed country effects (c1⋅⋅⋅c14), and a random disturbance at the context level (u0j).13 The latter is assumed to be normally, identically and independently distributed (??).14 Since the structure of the model is logistic, the binomial distribution of the vote intentions is assumed to adequately account for randomness at the individual level. All models were estimated in MLwiN 2.02 using the Penalized Quasi-Likelihood method based on a second-order Taylor expansion (PQL2).

Findings

Estimates for the four components of the model are presented in table 1. The rows in the lower third of the table contain the unit effects for the 14 countries under study, i.e. the logit of an ER vote when all individual and contextual variables are set to zero.15 While the coefficients themselves are of little intrinsic interest, their huge variation implies that even if a whole host of individual and contextual variables are controlled for, there are persistent differences between these countries that must be due to other, durable factors.

At the bottom of the table, σu02 represents the residual variance at the contextual level, i.e. the normally distributed random shocks that affect all voters in a given country at a given time. This figure is about 40 per cent lower than in a null model that contains only unit effects and a random term (not shown as a table), suggesting that the combination of contextual and individual variables goes a long way in understanding the puzzle of ER support. Nonetheless, a normal distribution with a variance of 0.3 will still produce a considerable number of rather large random shocks.16

The effects of the individual-level variables can be discerned from the topmost panel of table 1. As it turns out, the expected patterns re-appear even when one controls for contextual variables, unit effects and contextual variance. In line with theoretical expectations, groups who compete with immigrants for scarce resources and who have exhibited the highest level of xenophobia in the past – manual workers, younger voters, and the unemployed – show significantly more support for the ER than other groups. The gender gap is equally prominent. While the logistic link implies that effects are not linear-additive and that the proportion of ER support depends on the level of all independent variables, membership in either of these socio-demographic groups roughly doubles the probability of an ER vote.17 Again in line with previous findings, holding a university degree massively reduces the probability of a vote for an ERP, whereas being a pensioner has no significant effect on the vote.

Turning to the attitudinal variables, being a eurosceptic18 more than doubles the probability of an extreme right vote, but political dissatisfaction and ideology have even stronger effects. Dissatisfaction is operationalized through a four-point rating scale, therefore its maximal impact on the logit is 1.7 points. Left-right self-placement was measured on a ten-point rating scale, so its maximal effect on the logit is 4.7 points. Maximal effects paint a somewhat unrealistic picture since few voters hold extreme attitudes, but even if one considers the more conservative interquartile range of 1 point (dissatisfaction) and 3 points (ideology), it is obvious that political dissatisfaction and political leanings have significant and rather dramatic effects on the propensity to vote for the ER even when they are mutually controlled for.

Jointly, the coefficients in the upper panel of table 1 provide fresh evidence from a large number of political contexts that the ER vote is not based on protest alone, and that the ER is by no means a “catch all” that mobilizes all social groups in a similar way (Mayer and Perrineau 1992; see van der Brug and Fennema 2003 for a broader discussion). Rather, the ER’s success is based on its appeal to a constituency that has a distinct social and attitudinal profile.

This picture is complemented by th coefficients for the contextual variables, which are presented in the middle panel of table 1. “Disproportionality” and “decentralization” refer to the Gallagher-Index for the most recent election and the index devised by Lijphart (1999, 189), respectively. “Asylumseekers” reflects the number of new applications for asylum status per capita,19 “unemployment” refers to the standardized unemployment rates which are supplied by the OECD, whereas “unemployment benefits” reflects the impact of the OECD’s “Gross Unemployment Benefit Replacement Rates”.20 Three multiplicative interaction terms were created to reflect the hypotheses that the effects of unemployment and immigration reinforce each other (Golder2003b), while unemployment benefits can mitigate these effects (Swank and Betz2003).

“Toughness” (the most radical position on these issues taken by any party that is not considered to be part of the ER), “salience”, “variance” and the interaction of the latter two reflect information on party competition and political elite messages in a given context and were constructed as outlined above.21 To ease the interpretation and to reduce the likelihood of numerical problems, the unemployment, asylum and benefit rates as well as the measures for salience and variance were centered at their respective grand mean.

[Table 1 about here.]

In line with the findings by Arzheimer and Carter, decentralization and disproportionality of the electoral system do not have a statistically significant impact on support for the ER when other contextual variables are held constant. Again, the data provide no evidence that less proportional systems can curb support for the ER. It should, however, be borne in mind that (with the notable exception of France) the variation in the degree of proportionality is limited.22

The number of asylum seekers, unemployment rates,23 and the salience of the ER’s issues, however, all seem to have substantial and statistically significant effects on support for the ER. But in the presence of interaction terms, the size and statistical significance of these main effects must be calculated conditionally. The estimated effect for the number of asylum seekers, for instance, refers to a situation where both the centered unemployment rate and the centered level of unemployment benefits are held constant at a level of 0, i.e. at the mean of the original variables. Similarly, the positive effect of the unemployment rate is conditional on average levels of immigration and unemployment benefits.

[Figure 1 about here.]

Contrary to predictions derived from ethnic competition theory, and contrary to Golder’s (2003b) findings that are based on a different specification and on macro data alone, the interaction between levels of unemployment and immigration is negative: at higher-than-average levels of asylum applications and unemployment, the effects of both variables do not reinforce each other. Rather, a ceiling effect is observed that limits the impact of both contextual variables on the ER’s support. More specifically, if immigration is very high, unemployment does not matter any more. This is illustrated in Figure 1: at levels of immigration that are slightly higher than average (> 0.7), the effect of unemployment is not longer significantly different from zero (upper panel). Since the distribution of asylum application rates is right-skewed, this applies to roughly 20 per cent of all contexts. Note, however, that the effect of unemployment is rather weak even where immigration is at its empirical minimum of -0.98. Immigration, on the other hand, has a significantly positive effect even when unemployment is up to five points above its average (lower panel). This threshold is only exceeded in ten per cent of all contexts.

Similarly, more generous income replacement rates will reduce the effect of unemployment by a small and the effect of immigration by a considerable amount, as indicated by the two negative interaction effects. Moreover, unemployment benefits have an additional effect of their own, but this effect is small and statistically insignificant at average levels of immigration and unemployment. Even if unemployment and immigration rates are at their empirical extrema, this main effect will hardly effect the outcome.

[Figure 2 about here.]

The interpretation of the effects that political messages sent by other parties have on support for the ER is more straightforward. In line with Arzheimer and Carter, “toughness”, i.e. the ideological position of the most radical amongst the established parties, has no significant effect. This constitutes prima facie evidence against the “conspiracy of silence” hypothesis derived from spatial models of voting.

However, the two variables that reflect ideas of agenda setting and priming do have an effect. At an average level of variation in the party positions, a greater salience of the ER’s issues in the party manifestos is ceteris paribus related to a higher level of ER support. This effect (which prevails though objective factors like immigration and unemployment are controlled for) is quite pronounced: The interquartile range of the salience variable is 6.3, which translates into a change of 0.77 points on the logistic scale. This is roughly equivalent to the individual-level effect of being dissatisfied with European Integration.

The effect of salience is somewhat muted at higher levels of variation in the party statements as indicated by the negative sign of the interaction term, but even where the (centered) variance is higher than 50, the coefficient is significantly positive, as can bee seen from the lower panel of Figure 2. This applies to more than 95 per cent of all contexts. In fact, for 90 per cent of all contexts, the variance falls into the interval [-14.1; 36.6], implying that very often the negative interaction has no substantive consequences at all and the effect of salience prevails. The effect of variance, on the other hand, is neither statistically nor substantively significant, regardless of the level of salience.

[Figure 3 about here.]

While logit coefficients convey the direction of the effects and provide a means for testing their statistical significance, they are less useful for assessing the political relevance of a given variable. Here, the most relevant quantity is the predicted effect on the ER’s share of the votes, which depends on the respective level of all independent variables. A convenient tool for illustrating this impact are figures which plot the predicted probability as a function of one to three focal independent variables while all other independent variables are held constant at pre-specified levels that represent theoretically interesting “scenarios” (King, Tomz and Wittenberg2000). More recently, Mitchell and Chen (2005) have suggested that for complicated models, one could aggregate the average individual effects of all independent variables that are not varied in the graph into a single quantity which they dub “covariate contribution”. A relatively small number of covariate contributions (say three) could then be employed to cover a whole host of different “scenarios”.

This method is used in Figure 3 to illustrate the joint effect of unemployment rates, immigration, and unemployment benefits on the probability of a vote for the extreme right. Covariate components were obtained by calculating the logit for each of the 174,452 respondents from the fixed effects in the leftmost column of Table 1 and subtracting the joint impact of the three contextual variables and their interactions from this quantity. Then, the covariate contribution was set to the fifth, seventh, and ninth decile of its distribution. Levels of immigration and levels of unemployment benefits were set to the first, the fourth, the sixth, and to the ninth decile of their respective empirical distributions, while the unemployment rate was varied from its first to its ninth decile.

Figure 3 clearly demonstrates that while the effects of those three contextual variables that feature most prominently in previous research are statistically significant, their political relevance will often be negligible. The short-dashed lines that represent the lower half of all the individually calculated covariate contributions are basically flat, which indicates a trivial relationship between unemployment figures and the probability of a vote for the ER. Moreover, unemployment benefits and immigration levels hardly affect support for the ER, although weak effects are discernible if one compares the graphs within the same row or same column: support is minimal where benefits and immigration figures are close to their minimum (upper left corner), but increases very slightly (by less than two percentage points) where either of the two variables comes closer to its maximum.

If the contribution of other covariates is set to a somewhat higher level as represented by the solid lines, unemployment rates, unemployment benefits and immigration have a slightly stronger but rather complex effect on the predicted support for the extreme right. Comparing the graphs within each column reveals that higher levels of immigration are related to higher levels of ER support, although the differences are still small. Higher levels of benefits are related to higher levels of support, too, but this effect is restricted to contexts with below-average levels of immigration (the first two rows of graphs). As regards the effect of unemployment rates, a positive effect becomes visible, but only in contexts where either levels of immigration or unemployment benefits are very low. Most interestingly and in line with the findings by Swank and Betz, at high levels of immigration, unemployment benefits reduce the impact of unemployment, i.e. the line that represents this relationship is flat.

Finally, if the contribution of other covariates is set to a very high level (that could be due to individual-level effects, a strong unit-effect, the impact of other contextual variables or a large and positive random shock at the contextual level), the three contextual variables will have a strong and intertwining effect on the probability of a vote for the ER. Basically, a comparison within the columns of Figure 3 reveals that higher levels of immigration are associated with higher support for the extreme right. This effect, however, is much stronger where the level of welfare state protection is average or below average. At higher levels of unemployment benefits, the impact of immigration is much reduced. On the other hand, unemployment benefits are positively related to higher levels of support where immigration is low (c.f. the first two rows of Figure 3). Finally, unemployment figures are often strongly related to support for the ER, but where immigration is high, this relationship effectively vanishes, which reflects the lack of the positive interaction posited by Golder (2003b).

[Figure 4 about here.]

Analyzing the impact of party positions involves only two variables and their interaction and is therefore simpler. Figure 4 graphs the relationship between the salience of the ER’s issues in other parties’ manifestos and the expected vote share of the ER for four levels of variance in party statements and three levels of covariate contributions. As one would expect from the coefficients in Table 1 and the graphical analysis in Figure 2, the variance of party statements has very little impact on the success of the ER. The salience of the statements, however, is highly relevant for the ER’s electoral success, provided that the contribution of other covariates is large. For the lower 50 per cent of all covariate contributions, even very high levels of salience hardly increase the ER’s electoral fortunes, and at the seventh decile, the differences between high- and low-salience contexts are still small. If the contribution of other covariates falls into the upper third of its distribution, however, the political effect of salience is huge.

Summary

While the relationship between support for the ER and the contextual variables is much more complex than suggested by previous research, the basic results of the graphical analysis can be easily summarized. First, in line with theories of ethnic competition, the ER will benefit from high levels of immigration and unemployment, but this effect is moderated by the institutions of the welfare state. Generous unemployment benefits seem to curb the additional impact of unemployment where immigration levels are high. On the other hand, if immigration levels are very low, generous unemployment benefits increase the probability of an ER vote. Accordingly, the lowest levels of ER support are predicted for a system with minimal benefits, low unemployment rates, and minimal immigration. Extreme right mobilization would be most facilitated by high unemployment and high levels of either immigration or unemployment benefits (but not both). Independent of these objective social and economic conditions, political factors, i.e. the salience of the ER’s issues in the manifestos of other parties have a remarkable effect on the ER’s prospects.

Second, while these findings are of both political and theoretical interest, they apply only to a situation where the probability of an ER vote is already rather high due to other factors. For roughly 70 per cent of all covariate constellations, the contextual variables will have a small impact on the probability of a ER vote whereas for the remaining 30 per cent, contextual factors can tip the balance and can make an ER vote much more (or much less) likely.

Third, a considerable portion of the covariate contributions is due to country-specific intercepts and context-specific random effects. Consequentially, the political relevance of the contextual variables will be more pronounced (1) within sub-groups that exhibit disproportionate levels of support for the extreme right (e.g. politically dissatisfied right-leaning workers) but (2) also more generally in countries where the propensity of an ER is rather high across the board (Austria, Belgium, France, Denmark, Norway), and (3) in contexts that are affected by a substantial random shock (e.g. a media scare). The substantive implications of this latter possibility should not be underestimated: from the distributional assumption in Equation ?? and the parameter estimate of 0.3 in Table 1, it follows that roughly 35 per cent of all random shocks will shift the logit of an ER vote for all citizens in a given context by at least 0.5 points upwards or downwards. If a case is near the median of the covariate contributions, a difference of that size can render contextual variables politically relevant or irrelevant.

Conclusion

This paper set out from the twin question of why support for the ER varies so much across time and political systems. More specifically, its aim was to assess the impact of contextual variables on the support for the parties of the ER in Western Europe. The analyses presented here differ from previous accounts in two crucial ways:(1) The effects of individual and contextual variables are modeled jointly and(2) all relevant and available Eurobarometer data sets were included, resulting in maximal spatial and temporal coverage.

The findings on the individual level largely confirm previous results from national studies: The ER’s electorate has a clear social and attitudinal profile. These results rule out explanations that link the ER’s electoral appeals solely or chiefly on “protest”.

The picture at the contextual level is more complex. First, there is no empirical support for the “conspiracy of silence” hypothesis. On the contrary: in line with theories of agenda setting and priming, the salience of the ER’s issues (immigration and national identity) in the manifestos of the established parties has a strong positive impact, whereas the “toughness” of the established parties has no significant effect.

Second, both unemployment rates and immigration have generally a positive impact on the ER vote, but their respective effects do not reinforce each other. Rather, a ceiling effect is observed. Moreover, unemployment benefits can reduce support for the ER in certain constellations. Third, the political relevance of these effects crucially depends on the contribution of other covariates. Often, even constellations of contextual variables that clearly favor the ER will be of little political consequence.

Finally, even after differences in the composition of societies (via individual-level variables), features of the context, and random variation at the contextual level are taken into account, there are striking differences between countries as revealed by the estimates for the unit effects ranging from -8.7 to -3.2 points on the logistic scale. Put differently, given the levels of the variables included in the model, in Austria, Italy and Denmark the ER is persistently much stronger and in Spain, Sweden, and Finland, it is much weaker than one would expect it to be. Future research should focus on factors such as access to the media, organisational strength of the ER and links with other actors, political culture, and elite cues other than those in manifestos to come up with an explanation for these differences.

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PIC(a) Effect of unemployment
PIC(b) Effect of immigrationGraphs are based on estimates in Table 1. Both variables are centered. 95% confidence interval calculated with Variance-covariance matrix estimated under listwise deletion. The level of unemployment benefits is set to its mean of zero.

Figure 1: The conditional effects of unemployment and immigration

 


PIC(a) Effect of variance
PIC(b) Effect of salienceGraphs are based on estimates in Table 1. Both variables are centered. 95% confidence interval calculated with Variance-covariance matrix estimated under listwise deletion. The level of unemployment benefits is set to its mean of zero.

Figure 2: The conditional effects of salience and variance

 


PICGraphs are based on estimates in Table 1 (listwise deletion). All variables are centered. Covariate components set to -4.7 (short dash), -3.8 (solid line), and -2.3 (long dash).

Figure 3: The joint impact of unemployment, asylum seekers, and unemployment benefits on the probability of a vote for the extreme right

 


PICGraphs are based on estimates in Table 1. Both variables are centered. Covariate components set to -5 (short dash), -4 (solid line), and -2.7 (long dash).

Figure 4: The effect of the extreme right’s issues on the extreme right vote

 


Listwise Deletion Multiple Imputation
Male 0.482 (0.029) 0.485 (0.025)
18-29 years 0.437 (0.041) 0.419 (0.042)
30-45 years 0.194 (0.039) 0.172 (0.035)
>65 years -0.162 (0.054) -0.160 (0.047)
Education: middle/high 0.056 (0.036) 0.069 (0.034)
Education: university -0.324 (0.043) -0.251 (0.036)
Petty Bourgeoisie 0.043 (0.048) 0.088 (0.042)
Worker 0.370 (0.039) 0.350 (0.038)
Pensioner 0.054 (0.052) 0.034 (0.047)
Unemployed 0.471 (0.056) 0.484 (0.045)
Left-Right 0.552 (0.007) 0.505 (0.007)
Dissatisfied: EU 0.751 (0.036) 0.729 (0.038)
Dissatisfied: Democracy 0.607 (0.018) 0.551 (0.018)
Disproportionality 0.011 (0.016) 0.017 (0.016)
Decentralization 0.156 (0.158) 0.076 (0.162)
Asylumseekers 0.245 (0.056) 0.237 (0.057)
Unemployment 0.080 (0.032) 0.075 (0.033)
Asylumseekers ×Unemployment -0.024 (0.014) -0.031 (0.014)
Unemployment benefits 0.013 (0.010) 0.009 (0.010)
Unemployment benefits ×Unemployment -0.002 (0.002) -0.001 (0.002)
Unemployment benefits ×Asylumseekers -0.010 (0.005) -0.009 (0.005)
Toughness -0.038 (0.024) -0.033 (0.025)
Salience 0.122 (0.026) 0.128 (0.026)
Variance 0.008 (0.007) 0.006 (0.008)
Variance ×Salience -0.001 (0.000) -0.001 (0.000)
AT -3.271 (0.738) -2.963 (0.753)
BE -5.486 (0.665) -5.043 (0.674)
DE-E -7.060 (0.752) -6.624 (0.771)
DE-W -6.463 (0.764) -6.057 (0.783)
DK -4.990 (0.431) -4.700 (0.432)
ES -8.654 (0.520) -8.394 (0.591)
FI -7.370 (0.441) -6.934 (0.470)
FR -4.789 (0.354) -4.754 (0.361)
GR -5.564 (0.366) -5.284 (0.396)
IT -3.231 (0.336) -3.340 (0.351)
NL -7.444 (0.440) -7.097 (0.443)
NO -5.194 (0.437) -5.051 (0.445)
PT -6.272 (0.434) -5.781 (0.498)
SE -7.813 (0.600) -7.371 (0.598)
σu02 0.291 (0.033) 0.307 (0.037)
N(1) 174,452 267,348
N(2) 267 267

Logistic multi-level model. PQL2 estimates, model-based standard errors in parentheses. MI results based on eleven separate imputations.

Table 1: Support for the extreme right: Socio-demographics, attitudes, country effects, and contextual variables

*I wish to thank Paul Whiteley for advice and comments. I am also indebted to the participants of staff seminars at Essex and Mainz and to Elisabeth Carter, Jocelyn Evans and Chris Wendt for helpful comments at various stages of this research project.

1Much of the early literature is devoted to the perhaps not entirely fruitful twin debates on the “correct” label and on criteria for membership in this party family. However, at least the latter question is more or less settled since the mid-1990s: “we know who they are, even though we do not know exactly what they are” (Mudde 1996, 233, see also note 11). As regards terminology, this article refers to the “extreme right” because this seems to be the most commonly used label in recent research.

2On the other hand, theories of authoritarianism and anomia provide very limited analytical leverage because they focus on largely stable psychological states. Therefore, it is difficult to see how they could explain short-term fluctuations of ER support within a given country or persistent differences between otherwise largely similar countries.

3Jackman and Volpert analyze 103 elections that were held in 16 countries between 1970 and 1990.

4The relatively small Christian Social Union in Germany is a point in case: while they are clearly positioned in the political mainstream and have always formed an alliance with the much larger Christian Democratic Union at the national level, they take a tougher stand on immigration than the “Post-Fascist” Italian National Alliance (Lubbers, Gijsberts and Scheepers2002).

5According to Zaller, this effect is moderated by the political awareness of the respondents. In principle, the role of political awareness could be modeled by a cross-level interaction. However, since data on political awareness is rather limited, and since this moderating effect is not central to the argument presented in this paper, this route was not pursued.

6Switzerland is excluded from the analysis both for substantial reasons as well as for a lack of data.

7The partial cumulation of the Eurobarometer produced by a team led by Hermann Schmitt (Schmitt et al.2002) greatly facilitated the construction of the data set. An appendix containing details of the coding and imputation procedures as well as scripts for Stata and Mlwin that can be used to replicate the findings, additional tables and an assessment of the robustness of the findings are available through the author’s dataverse at .

8Because of the economic, social, and political-cultural differences, there are separate surveys for East and West Germany. Norway did not accede the EU in 1994 but did participate in the Eurobarometer between 1990 and 1996 and then again in 2002/2003.

9Luxembourg had to be excluded because the OECD does not calculate standardized benefit rates for this country. Estimates for a number of more parsimonious models that are based on a larger number of respondents and contexts including Luxembourg are presented in the online appendix.

10Britain is a case in point. In 2005, the British National Party, now the most important party of the ER in the United Kingdom (Eatwell2004), contested 119 of the 646 Westminster constituencies, i.e. less than 20 per cent (Norris and Wlezien2005, 678). With 57 and 33 candidates, the numbers were even lower in 1997 and 2001 (Yonwin2004, 7).

11The variable is coded as 1 if an respondent intends to vote for the Freedom Party in Austria, the Front National or the Vlaams Blok in Belgium, the Freedom Party or the Danish People’s Party in Denmark, the Rural Party or the True Fins in Finland, the National Front in France, the German People’s Union, Republikaner or National Democrats in Germany, the EPEN, the National Front and Political Spring in Greece, the National Alliance and the Northern League in Italy, the Center Parties and the Lijst Pim Fortuyn/Leefbar Nederland in the Netherlands, the Freedom Party in Norway, the “Christian Democrats” in Portugal, the various Falange Parties in Spain, and New Democracy in Sweden. Voters of other parties and self-declared non-voters are coded as 0.

12To simplify the presentation, the model contains no constant but rather one unit-dummy for each country.

13Note that a double index indicates variation both across persons and contexts, while variables with a single index vary across contexts but are constant over persons within the same context.

14More explicitly, the Variance-Covariance Matrix Ωu that governs the distribution of u0j is assumed to be a diagonal matrix whose elements are identical.

15AT = Austria, BE = Belgium, DE-E = East Germany, DE-W = West Germany, DK = Denmark, ES = Spain, FI = Finland, FR = France, GR = Greece, IT = Italy, NL = Netherlands, NO = Norway, PT = Portugal, SE = Sweden.

16If one considers a shock of 0.7 points on the logistic scale (which is equivalent to the effect of Euroscepticism) as “large”, this threshold will be exceeded in about 20 per cent of all realizations.

17All socio-demographic variables enter the model as dummy indicators.

18Euroscepticism is measured by a dummy variable.

19Other figures such as the number of non-white residents or the share of foreign-born residents could have been employed, too, but the data on asylum seekers and refugees are preferable for at least three reasons: first, asylum applications (and family reunification claims, which are often related to previous applications) have provided the main route for new legal immigration into Wester Europe since the 1970s (Freeman1998, 94), second, unlike other measures they are comparable across time and countries, and third, asylumseekers and refugees have become the main focus of the ER’s propaganda. To prevent numerical problems, the numbers were entered as applications per 1,000 residents.

20The OECD calculates “Gross Unemployment Benefit Replacement Rates” by averaging over several types of households, durations of unemployment, and income levels before unemployment.

21Parties of the ER were excluded from all calculations. For salience and variance, the figures were weighted according to the relative size (vote share) of the respective parties. Since the party manifestos are usually published only when an election is imminent, party positions between publication dates were interpolated.

22Removing France from the sample does not substantively affect the results.

23The individual employment is controlled for.

 

Christian Religiosity and Voting for West European Radical Right Parties

 

The academic literature on parties and voters of the extreme, radical or populist right is vast, and from this work we know that some voters are more likely than others to vote for these parties. The effects of certain socio-demographic characteristics on the radical right vote have been very well documented and there is a consensus in this literature that male voters, young voters, voters with low or middle levels of education and voters from certain social classes are more likely to vote for radical right parties than are other electors (see for example Arzheimer and Carter 2006; Betz 1994; Lubbers et al. 2002). Studies also agree that the attitudes of voters impact on their likelihood of casting a vote for these parties and that negative attitudes towards immigrants are particularly powerful in predicting a vote for a radical right party (Billiet and De Witte 1995; Lubbers et al. 2002; van der Brug et al. 2000).

 

Within this body of literature the impact of a voter’s religious attachment, involvement and attitudes on his or her propensity to vote for a party of the radical right has received relatively little attention, at least as compared to the effects of gender, age, education or class and the influence of certain attitudes. This is not wholly surprising given the importance of these other predictors. Furthermore, models of radical right voting are likely to have omitted variables that relate to religion for practical reasons: reliable, comparative data on religious behaviours and beliefs are hard to come by.

 

We believe, however, that there are valuable reasons for investigating the link between a voter’s religious attachments and beliefs and his or her likelihood of voting for a radical right party. And this is not because of the ever-present academic desire to ‘fill a gap in the literature’, although a gap does clearly exist (Mudde 2007: 296). Rather, in the first instance, our desire to explore this relationship rests on the widespread acknowledgement that, despite their decline (Crewe 1983; Crewe and Särlvik 1983; Dalton et al. 1984), traditional social cleavages continue to be important in structuring partisan alignments and electoral choice (Mair et al. 2004), and that the divide between religious and secular voters is still a relatively strong predictor of vote (Dalton 1996). To begin with, therefore, we are guided by research such as Girvin’s, which argues that ‘although electoral behaviour is affected by other factors such as gender and class, church attendance in a number of cases is the single most important variable in explaining voting decisions’ (Girvin 2000: 13; see also Norris and Inglehart 2004).

 

Secondly, we would argue that it is useful to concentrate on the impact of religion on a specific electoral choice – namely the likelihood of a vote for the radical right – because such a focus will ultimately tell us more about the role of religiosity in electoral choice. As we shall see, there are a number of good reasons to suggest that religiosity will reduce the likelihood of a vote for the radical right, and yet there are also good reasons to suggest that it might increase this likelihood. By disentangling the various influences of religiosity on the radical right vote, and by assessing their strength, we may gain a better understanding of the ways in which religiosity does or does not affect electoral choice in general.

 

In this article we therefore propose to investigate the impact of religiosity on the radical right vote because this endeavour serves a dual purpose: from the religiosity end of the telescope we seek to learn more about the impact of religiosity on electoral choice, while from the radical right end of it, we aim to gain an understanding of the predictive strength of religiosity on the radical right vote.

 

It also transpires that we have chosen to point our telescope into the sky at a rather interesting time. To be sure, traditional social cleavages have weakened and levels of church membership and religious participation have declined (Girvin 2000), yet religion has also rather unexpectedly assumed a greater centrality in the political life of West European societies in recent years. Its return to the global political agenda – as evidenced most pronouncedly by the war between Al Qaeda and ‘the West’ – has had considerable domestic implications in Western Europe, aggravating tensions between Christian or agnostic majorities and a host of minority groups that are increasingly defined (by themselves and the outside world) not in ethnic, but in religious terms. Conflicts over the symbolism of headscarves worn in public institutions in France, rows about veils in the UK, death-threats aimed at female politicians from Islamic backgrounds in the Netherlands and in Germany, and the crisis over the Danish cartoons are just some examples of such tensions. While it is too early to gauge the precise impact of such developments on long-term electoral choices, this context does make our decision to revisit the link between religiosity and electoral choice rather timely.

 

The rest of this article follows a conventional structure: the next section outlines our conceptualization of religiosity and our favoured terminology, and sets out our theoretical framework and hypotheses. We then explain our model and our variables, and describe our data and methodology. Having done this, we present our results and discuss our findings. We close with an assessment of the importance of religiosity in predicting electoral choice both for radical right parties and indeed more generally.

 

 

Religiosity and voting for the radical right: conceptualization and theoretical framework

 

As mentioned above, few studies have explored the impact of a voter’s religious attachment, involvement and attitudes on his or her likelihood of voting for a party of the radical right. What is more, those that have devoted attention to this question have, in the main, been single-country studies (e.g. Billiet 1995; Billiet and De Witte 1995; Lubbers and Scheepers 2000; Mayer 1998; Mayer and Perrineau 1992; van der Brug 2003; Westle and Niedermayer 1992). There are just four cross-national studies of radical right voting that have included an examination of the effect of religiosity, and the findings of these were rather mixed in that two found that religiosity had weak and inconsistent effects on party preference (van der Brug et al. 2000; van der Brug and Fennema 2003), while the other two concluded and that less religious (Norris 2005: 138-9) or non-religious (Lubbers et al. 2002: 348) people were over-represented in the radical right electorate.

 

Crucially, and in stark contrast to the more recent studies that examine the relationship between church involvement and ethnocentrism or prejudice (e.g. Billiet et al. 1995; Eisinga et al. 1990, 1999), these comparative analyses conceptualize and operationalize religiosity in a rather simple way: van der Brug et al. (2000) and van der Brug and Fennema (2003) include a composite variable in their models, which is made up of religious denomination and church attendance, Norris (2005) makes use of a measure of religious self-identification, and Lubbers et al. (2002) distinguish between non-religious people, religious people belonging to non-Christian denominations, and Christian people. We would argue that these conceptualizations and operationalizations are problematic because they are too blunt to untangle the different effects that religiosity may have on the likelihood of radical right vote and, as a result, they are likely to underestimate the total effect of religiosity (Bartle 1998). Research on religiosity and ethnocentrism (discussed below) suggests that religious affiliation, involvement and belief structures can be linked to the radical right vote in different ways and so it is crucial to conceptualize religiosity in a manner that captures its different aspects or dimensions, and the ways in which these might interact. To this end we conceptualize religiosity as a combination of religious affiliation, church attendance, private religious practice and self-stated religiosity. Precisely because our conceptualization captures the different aspects of religious activity and beliefs, we favour the term ‘religiosity’ over ‘religiousness’ or simply ‘religion’.

 

As indicated in the introduction, there are reasons to believe that religiosity may reduce the likelihood of a radical right vote, and yet there are also reasons to believe it may increase it. Focusing first on why religiosity might reduce the likelihood of such a vote, to begin with there is plenty of evidence to suggest that religious affiliation and involvement will lead to a greater likelihood of a voter voting for a party of the mainstream right, such as a Christian, Christian Democratic or conservative party that has traditionally defended religious interests, than any other type of party, including a party of the radical right. Of course Christian and Christian Democratic parties differ from conservative parties in terms of their origins and ideologies, with the former traditionally defending Christian values and the latter having no links with organized religion, but both long-standing research and more contemporary studies have shown that religious voters have tended to favour parties of the mainstream right, irrespective of whether these parties are of the Christian Democratic or the conservative type.

 

Many analyses of voting in Weimar Germany report that the Catholic electorate was less permeable to the NSDAP than other sections of society, and attribute this to Catholic voters’ attachment to the Zentrum party, as well as to the integrating role played by Catholic networks and organizations (e.g. Childers 1983: 188-9; Falter 1991; Grunberger 1971: 552; Lipset 1971: 147-9; Mommsen 1996: 353). And despite widespread secularization, the attachment of religious voters to Christian Democratic or conservative parties continues to be observed today. Norris and Inglehart, for example, argue that ‘in industrial and postindustrial societies […] religious participation remains a significant positive predictor of Right orientations’, even after controlling for a whole range of other socio-demographic, economic and contextual factors. Indeed, they conclude that ‘religious participation emerges as the single strongest predictor of Right ideology in the model, showing far more impact than any of the indicators of social class’ (2004: 204-7. See also Girvin 2000: 21). Given these findings, we believe it is therefore reasonable to expect a certain degree of ‘encapsulation’ of religious voters by Christian, Christian Democratic or conservative parties (see Hypothesis H1 below).

 

Secondly, we also expect religious voters to be less likely to vote for a party of the radical right than other voters for the simple reason that radical right parties will not appeal to them (see Hypothesis H2a below). On the one hand, radical right parties do nothing to attract religious voters since they do not discuss religion in their ideologies and programmes. Instead, these parties have only addressed the subject for purposes of political advantage and mobilization and/or because it fits in with their world-view. For example, the parties are much more concerned about non-Western religions (particularly Islam) that are said to be a threat to Western culture and society than they are about any of the moral substance of religious teachings, or about what adhering to a faith might actually mean and entail. In some specific cases the radical right’s failure to appeal to religious voters is also explained by anti-clerical traditions (as in Austria and Germany), or by the fact that the parties have libertarian roots (like in Norway and Denmark). On the other hand, the issues that the parties do discuss and the views they have on these issues are often very much at odds with the beliefs and values of religious voters. After all, the values, beliefs, and traditions associated with most contemporary versions of the Christian faith are those of tolerance, compassion and altruism, and these find little in common with the authoritarian, xenophobic and even racist ideologies and appeals of the parties of the radical right, and the practice of targeting some of the most vulnerable groups in society such as refugees and immigrants.

 

For a number of different reasons, therefore, it is wholly reasonable to suggest that religiosity might ‘insulate’ voters from the appeals of a party of the radical right. However, for a variety of other reasons, it also makes sense to hypothesize the contrary, and to expect religious affiliation, religious involvement and the intensity of religious beliefs to be linked with a greater support for a party of the radical right. As regards affiliation, a number of studies, starting with that by Allport and Kramer (1946), have concluded that people with no religious affiliation show lower levels of ethnocentrism than people who describe themselves as Catholic or Protestant (see also Pettigrew 1959). As for religious involvement, dozens of analyses have pointed to the existence of a relationship between church attendance and levels of prejudice. The seminal work by Adorno et al. (1950) was one of the first to report a curvilinear relationship between church attendance and prejudice. While, in general, it found higher levels of ethnocentrism among churchgoers than among non-attenders, more specifically it found that regular churchgoers and non-attenders were both less prejudiced than those who attended church on a less frequent or an irregular basis. A number of subsequent analyses, carried out both in the US and in Europe, have reached similar conclusions (e.g. Allport and Ross 1967; Eisinga et al. 1990; Gorsuch and Aleshire 1974; Petersen and Takayama 1984; Pettigrew 1959; Studlar 1978). Other studies have proposed that prejudice also depends on the nature of particular religious convictions or belief structures and that people with strong religious beliefs are prone to developing a ‘closed belief-system’, which has often been linked to ethnocentrism and authoritarianism (Glock and Stark 1966; Rokeach 1960; but see also Middleton 1973; Ploch 1974; Roof 1974 for a critique of this argument). While many of the studies just mentioned may reflect a climate specific to the United States of the 1950s and 1960s, a link between closed religious belief-systems and ethnocentrism has also been uncovered in a more recent analysis of religion and prejudice (Altemeyer 2003) as well as in a recent pan-European youth survey (Ziebertz et al. forthcoming).

 

To be sure, many of the early studies on religiosity and prejudice have been criticized on theoretical, conceptual and methodological grounds (see Eisinga et al. 1999 for a useful summary). Many failed to ascertain whether religious doctrines act as a trigger for prejudice, or whether, conversely, they legitimate existing prejudices. In addition, these early studies have been attacked for failing to adequately specify both dependent and independent variables, and in particular for muddling up different dimensions or aspects of religiosity, such as affiliation, church attendance, and belief structures (Scheepers et al. 2002). Finally, many of these early works also tended to examine bivariate relationships only, and did not control for other social variables such as age, educational level, class, or localism.

 

Despite the shortcomings of these studies, however, there is still good reason to hypothesize that religiosity may be linked with a greater propensity to vote for a radical right party because the literature cited above clearly points to a link between religiosity and ethnocentrism. And since negative attitudes towards immigrants – which are closely related to ethnocentrism – are one of the most powerful predictors of a vote for a party of the radical right (as discussed above), it makes sense to hypothesize a two-step link between religiosity, anti-immigrant sentiment and voting for a radical right party, with religious people showing a greater likelihood of voting for the radical right than other people (see Hypothesis H2b below).

 

On the basis of these arguments, a number of hypotheses – which bring together different strands of theory that have not been considered in combination before – may be advanced as to the impact of religiosity on the likelihood of a radical right vote. Of course, despite these arguments, it could well be that religiosity is not a cause of radical right thinking, but is instead a correlate, since religious people are not only older (Argue et al. 1999), but also tend to have lower levels of education (see Johnson 1997) and therefore are less likely to embrace liberal-democratic values than their compatriots. We therefore also advance a hypothesis that proposes that religiosity has no direct effect on the likelihood of a radical right vote, and that instead, any effect is due to socio-demographic characteristics alone (Hypothesis H3 below). Our (competing) hypotheses are as follows:

  •  H1: Religious people are less likely to vote for the radical right because they are firmly attached to Christian Democratic or conservative parties;
  • H2a: Religious people are less likely to vote for the radical right because they are less likely to adopt negative attitudes towards immigrants;
  • H2b: Religious people are more likely to vote for the radical right because they are more likely to adopt negative attitudes towards immigrants;
  • H3: All direct relationships between religiosity and the vote are spurious (i.e. once radical right-wing attitudes and party identification are controlled for, the remaining effects of religiosity are due to the socio-demographic profile of religious people and will disappear completely if group memberships are taken into consideration.)

 

In principle, these mechanisms can reinforce or counterbalance each other. In addition, the extent to which these hypotheses may be borne out in practice will clearly depend on differences in national contexts and on features of each political system. It is well beyond the scope of this study to examine these differing national contexts (see for example Broughton and ten Napel 2000; Hanley 1994; van Hecke and Gerard 2004), but as a starting point we may point to the importance of differences in the strength of the religious cleavage. In the Lutheran countries of Scandinavia the religious cleavage is relatively weak (Madeley 2004), and so encapsulation by Christian Democratic parties is likely to be moderate at best. By contrast, in denominationally mixed countries (such as the Netherlands and Switzerland), where this cleavage is stronger, greater encapsulation is to be expected. Secondly, any traditional links between the church and specific political forces are likely to be relevant. In France, for instance, there has historically been a close connection between fundamentalist streams within the Catholic Church and anti-modern and illiberal political forces (Minkenberg 2003; Veugelers 2000). In this context, religiosity is likely to have a quite different connotation than in countries that lack such a tradition.

 

The characteristics of individual parties will also have an effect on our findings. Most obvious is whether the parties of the mainstream right are Christian Democratic or, as in France, are conservative parties. Even where Christian Democracy prevails significant differences exist between the parties: while some parties, such as the Austrian ÖVP, are catch-all parties that have attempted to integrate a host of different ideological tendencies (Fallend 2004), others, like the Belgian CVP and PSC remain confessional parties (Lucardie and ten Napel 1994). Different still are the Scandinavian Christian Democratic parties, which emerged much later and which grew ‘out of traditions of religious dissent representing various shades of dissatisfaction with the religious establishment among activist minorities’ (Madeley 2004: 218). On a more specific, policy-level, some Christian Democratic parties have tended to stress the Christian values of compassion and tolerance and are therefore inclined to support the rights of immigrants (see della Porta 2002 on the case of Italy, where a strong, Catholic pro-immigrant movement exists), whereas others– like the German CSU– have taken a tough stand on immigration (Lubbers et al. 2002: 356).

 

Radical right parties also differ in their ideological profiles (Betz 1994; Carter 2005; Ignazi 1992; Kitschelt 1995; Taggart 1995) and these differences are likely to have implications for our findings since the parties will attract different socio-economic segments of the electorate, and will entice voters with different attitudes. While most parties of the radical right have no specific interest in religion, the French Front National has always (not least through its stand on abortion) tried to appeal to conservative Catholics, and the Italian Alleanza Nazionale is actively trying to develop a more Christian/conservative profile. The latter party is also unusual insofar as it places much less emphasis on the issue of immigration and is much less xenophobic than most other parties of the radical right (della Porta 2002). For all these reasons, therefore, we certainly expect country differences. That said, it is not our intention here (especially with only eight cases) to test explanations for these differences, even though we can engage in some speculation as regards our results.

 

 

Modelling the links between background variables, religiosity and the radical right vote

 

Although our model is a little complicated, its basic structure (see Figure 1) is of the simple block-recursive type that has been fruitfully applied in electoral research before (Bartle 1998; Miller and Shanks 1996) and that helps us establish the direction of the flow of causality. Located at the very beginning of the causal chain are several socio-demographic variables that are exogenous: although these socio-demographics will often affect the level of religiosity as well as the development of political attitudes and the vote, it is inconceivable that religiosity will cause gender, age, education or class. Religiosity in turn can have a causal effect both on political attitudes and on behaviour, but it is implausible to assume the reverse. Finally, the vote itself depends (amongst other things) on attitudes, religiosity and socio-demographic features but does itself not alter these variables.

 

In contrast to the comparative studies mentioned above, which included religiosity as an independent variable, our model incorporates religiosity as a variable that appears before political attitudes in the causal chain. This allows us to consider the different ways in which religiosity may affect the likelihood of a radical right vote. In particular we can examine whether its effects are direct, indirect, or are due to background variables (i.e. whether they are spurious).

 

[FIGURE 1 ABOUT HERE]

 

The actual model on which our analysis is based is represented in Figure 2. The dependent variable in the analysis is vote for a party of the radical right, as depicted on the right hand side of the diagram (Block IV). This, we argue, is likely to be influenced by three sets of independent variables: religiosity (Block II); radical right attitudes (Block III); and socio-demographics (Block I). In addition, it is likely to be influenced by an intervening variable, namely an individual’s party identification with a Christian Democratic or conservative party (labelled ‘CD-PID’). This is also located in Block III.

 

[FIGURE 2 ABOUT HERE]

 

We begin by considering the impact of the three sets of independent variables independently of each other. The variable ‘Religiosity’ is a latent variable constructed from four observable variables (rel1-rel4) that tap the different aspects of religiosity that previous research has identified, namely religious affiliation, church attendance, private religious practice and self-stated religiosity (see below for further details on the data). We treat these variables as indicators of a single latent variable because they are highly correlated in all countries under study. This allows us to deal with one variable only and yet to continue to benefit from the advantages that multi-indicator variables bring in terms of enhanced reliability and validity of results. As alluded to above in Hypothesis 3, independent of any identification with conservative or Christian Democratic parties and independent of an individual’s radical right attitudes we expect to see no direct relationship between religiosity and the radical right vote because the parties of the radical right pay little attention to religious issues.

 

The early studies discussed above examined the link between religiosity and ethnocentrism – i.e. a tendency to regard one’s own ethnic and cultural group as superior and to treat other groups with contempt (Sumner, 1906). We would argue that, since (non-Western) immigrants make up the most prominent ‘out-group’ in West European societies, it makes sense to operationalize this concept by including variables that capture an individual’s attitudes towards immigrants. ‘Radical Right Attitudes’ are therefore measured by 21 observable attitudinal variables (labelled rra1, rra2 etc in Figure 2) that relate to views on immigrants and refugees. Empirically, these 21 variables show a very high degree of intercorrelation and are thus treated as indicators of a single latent variable. Clearly, since previous research has shown that anti-immigrant sentiment is one of the strongest predictors of a radical right vote, we expect to see a positive relationship between this variable and the radical right vote.

 

Our third set of independent variables is composed of socio-demographic variables. These include age, gender, class and education. In line with the findings of previous studies, we expect a greater propensity to vote for the radical right among younger voters as compared to older voters, among male voters as compared to female voters, among voters with lower levels of education compared to those with high levels of education; and among working-class voters, farmers and the ‘petty bourgeoisie’.

 

As regards our intervening variable (‘CD-PID’) that refers to voters’ identification with a Christian Democratic or conservative party, clearly, we expect voters who identify with such parties to be less likely to vote for a party of the radical right than voters who display no such identification.

 

Of course, the three independent variables just discussed are not expected to exert an effect on the propensity of a radical right vote in isolation only. Rather, socio-demographic variables are likely to have an impact on an individual’s religiosity, and on his or her attitudes. This is shown in Figure 2 by arrows that flow from ‘Socio-Demographics’ to ‘Religiosity’, and from ‘Socio-Demographics’ to ‘Radical Right Attitudes’. In addition, socio-demographics are likely to have an impact on the likelihood of an individual’s identification with a Christian Democratic or conservative party, hence the further arrow that runs from ‘Socio-Demographics’ to ‘CD-PID’. We also cannot rule out the possibility that the socio-demographics have a direct impact on the vote after controlling for religiosity, radical right attitudes, and ‘CD-PID’, and there is therefore an arrow connecting ‘Socio-Demographics’ and ‘Radical Right Vote’ directly, capturing any residual effects of group membership on the vote that might remain after controlling for attitudes. These include any spurious effects of religiosity (Hypothesis H3).

 

Religiosity, for the theoretical reasons discussed above, is likely to have either a negative or a positive impact on radical right attitudes (Hypotheses H2a and H2b). This is shown by the arrow in Figure 2 that runs from ‘Religiosity’ to ‘Radical Right Attitudes’. In addition, we expect religiosity to have an effect on identification with a Christian Democratic or conservative party.

 

Radical right attitudes are very likely to have a direct effect on the vote for the radical right. Yet we cannot rule out that they might additionally be correlated with ‘CD-PID’ because people who identify with established, mainstream right-wing parties may be more likely to hold radical right attitudes than other citizens. That said, we can make no assumption as to the direction of this relationship, and so our model depicts a mere correlation, as represented by a double-headed arrow running between ‘Radical Right Attitudes’ and ‘CD-PID’.

 

This model enables us to test whether religiosity influences the radical right vote in any way whatsoever. If religiosity does affect the radical right vote, the model allows us to test whether it does so directly, or indirectly (through radical right attitudes and/or an identification with a Christian Democratic or conservative party), or whether the effect of religiosity is spurious (i.e. related to socio-background variables). The model thus allows us to test a number of alternative ‘routes’ that have so far largely been neglected or conflated in the literature on religiosity and on the radical right.

 

 

Data and Methodology

Our data come from the first round of the European Social Survey (EES), the fieldwork of which was conducted in 2002. This database is particularly attractive because it includes a whole host of measures of radical right attitudes as well as of religious views and behaviours. From the 22 countries covered in this survey we selected eight West European systems that have witnessed a substantial and persistent support for the radical right: Austria, Belgium, Denmark, France, Italy, Netherlands, Norway and Switzerland. While countries in which the radical right has been unsuccessful should be included in macro-level explanations of party success so as to avoid selection bias, it makes no sense to include them in micro-level models. If not a single respondent reports the intention to vote for the radical right (as in Spain, Sweden, or the UK), there is simply nothing to model. By much the same token we excluded Germany as the number of self-declared radical right voters here was tiny (n=10), making conventional logit or probit modelling unfeasible.

 

Respondents under the age of 18, non-citizens, and members of non-Christian faiths were excluded. In six of the eight countries included in this study there was little variation in the denomination of respondents who indicated they were of a Christian faith. Only in the Netherlands and Switzerland were there significant numbers of both Catholics and Protestants. The impact of different religious doctrines can therefore only be examined in these two countries, and this is confined to noting differences between Catholic and Protestant voters only, since the ESS does not disaggregate between different strands of Protestantism.

 

All respondents who stated that, in the last election, they had voted for the Austrian Freiheitliche Partei (FPÖ), the Flemish Vlaams Blok (VB) or the Belgian Front National (FNb), the Danish Dansk Folkeparti (DF) or Fremskridtspartiet (FRPd), the French Front National (FN) or Mouvement National Républicain (MNR), the Italian Alleanza Nazionale (AN), Lega Nord (LN) or Movimento Sociale-Fiamma Tricolore (Ms-Ft), the Dutch Lijst Pim Fortuyn (LPF), the Norwegian Fremskrittspartiet (FRPn), or the Swiss Freiheitspartei der Schweiz (FPS), Lega dei Ticinesi (LdT), Schweizer Demokraten (SD) or Schweizerische Volkspartei (SVP) were given a code of 1. All remaining respondents were given a code of 0. There was an average of 1,700 respondents per country.

 

As regards the socio-demographic variables we coded male respondents as 1 and female respondents as 0, and we recoded age into three categories that reflect the findings of previous studies on its effects on the radical right vote (18-29; 30-65; older than 65). For social class, data was first mapped onto the familiar Goldthorpe-Scheme. Then, to keep things as simple as possible, we created a dummy variable that takes the value 1 for those classes that have shown the greatest support for the radical right in the past – workers, farmers, and the petty bourgeoisie – and 0 for all others. For education we used the ESS’s seven-point scale of achievement that ranges from ‘no primary education’ (1) to ‘second stage of tertiary education’ (7).

 

We made use of the four measures contained in the ESS that capture different aspects of religious activity and beliefs. The first two concern the regularity with which an individual prays outside of religious services and the regularity with which he or she attends religious services (other than on occasions such as weddings, funerals etc.). These were each measured on a seven-point scale ranging from 1 (‘every day’) to 7 (‘never’). We reversed both scales to facilitate interpretation. The third measure taps religious affiliation and simply asks whether the respondent belongs to a Christian church or considers him or herself to be a Christian. Respondents who replied in the affirmative were coded as 1 and all others were coded as 0. The final measure of religiosity asks the respondent for a self-assessment of religiosity and is measured on a scale that ranges from 0 (‘not at all religious’) to 10 (‘very religious’). As is clear from its wording, this question is not about formal religious membership. It can thus be interpreted as a measure of the intensity of non-institutionalized Christian beliefs.

 

Identification with a Christian Democratic or conservative party in the sense of the Ann-Arbor model was operationalized as a simply dummy variable. Respondents who identified with the ÖVP in Austria; the CVP (now CD&V) or PSC (now CDH) in Belgium; the KF or KD in Denmark; the RPF, UMP or UDF in France; the CCD-CDU (now UDC), Forza Italia or NPSI in Italy; the CDA, CU or SGP in the Netherlands; the KRF or Høyre in Norway; and the CVP or EVP in Switzerland were coded as 1, while all others were coded as 0.

 

Finally, as mentioned above, we selected 21 observable attitudinal variables from the ESS to construct our latent variable ‘Radical Right Attitudes’. These cover a number of subdimensions of radical rightist thinking including attitudes towards the economic, social and cultural impact of immigrants, attitudes towards race and ethnicity, and attitudes towards immigrant and refugee rights. These variables were measured on a variety of scales. The full details of all 21 variables, as well as the full datasets for each country, can be found in the replication archive at http://hdl.handle.net/1902.1/12312

 

Since we have a significant number of variables in our model we did not use listwise deletion. Rather, we employed Multiple Imputation by Chained Equations (MICE), a very versatile imputation method that fills the gaps in the data set with a range of ‘plausible’ values. As our core dependent variable, one intervening variable and several of our indicator variables are dichotomous, we estimated the models with an extension of the Structural Equation Modelling (SEM) framework, implemented through the program MPlus, which allows for transparent handling of categorical variables (see Muthén 2004 for an overview).

 

To identify our model, the scales of the two latent variables (religiosity and radical rightist attitudes) had to be fixed. We did this by setting the coefficients for the paths from the latent constructs to an arbitrary indicator (praying and wages respectively) to one. Since we expect the basic structure outlined in our model to apply in all countries but the actual strength of the relationships to vary across systems, we estimated our models on a per-country basis with no equality constraints. Most parameters presented in the tables below are unstandardized regression coefficients. Exceptions are the effects on the dichotomous variables (identification with a Christian Democratic or conservative party, belonging to a Christian church/considering oneself a Christian, and radical right vote), which are represented by unstandardized probit coefficients. While all the relationships between variables were estimated simultaneously, we will discuss our findings from each regression in turn, so as to make interpretation easier.

 

 

Religiosity and radical right voting: findings and discussion

The overall fit between our model and our data is good. The Root Mean Square Error of Approximation is well below the conventional threshold of 0.1 in all countries and comes close to 0.05 in most countries, which indicates a ‘very good’ fit. The measurement models for religiosity and radical right attitudes also perform very well: all coefficients are significant (throughout this article we use the conventional 5 per cent threshold) and positive. Moreover, all are, by and large, within the same range. Full details of these measurement models can be found at http://hdl.handle.net/1902.1/12312.

 

Turning now to the substantial relationships, Table 1 shows the regression of religiosity on the socio-demographics and enables us to see which of the different groups in the eight societies are, on average, more (or less) religious. The findings again point to a largely uniform pattern across the countries: holding other socio-demographic variables constant, men are considerably less religious than women and older citizens are more religious than younger people. Importantly, since the age groups 30-65 and 66+ have large positive coefficients, Table 1 also indicates that young men – who make up the social group that shows a disproportionally high level of support for the radical right in all West European countries – are also the group least likely to be religious. By contrast to gender and age, education (with the exception of Switzerland and Italy) and class have no significant effects once the other variables are controlled for.

 

[TABLE 1 ABOUT HERE]

 

Next, since previous research has shown that radical right-wing attitudes are an excellent predictor of the radical right vote, we turn our attention to the antecedents of these attitudes. As can be seen in Table 2, we find that education has a significant and strong negative effect on radical-right attitudes in all eight societies under study even when the other socio-demographic variables and religiosity are held constant. This result is in line with existing research that found that higher levels of education are usually associated with more liberal views (Coenders and Scheepers 2003; Weakliem 2002). Class has the expected significant positive effect on radical-right attitudes: working-class voters, farmers and voters categorized as belonging to the petty bourgeoisie show a greater propensity of holding radical right-wing attitudes than other class groups even after controlling for education. The only exception here is the Netherlands, where the effect of class is still positive but is somewhat weaker and is not statistically significant. The effect of age on radical right-wing attitudes is mostly positive – i.e. older people have, on average, and after controlling for the other factors, slightly more radical right-wing attitudes than their younger compatriots. The two exceptions here are Italy, where age effects are reversed, and the Netherlands, where they are insignificant. By contrast, gender has no discernible effect on radical right-wing attitudes, with the exception of Norway, where men have somewhat more radical right-wing attitudes than women.

 

Finally, with respect to religiosity, we find that this variable has hardly any effect at all on people’s attitudes towards radical right issues: in five of the eight countries (including the two denominationally mixed ones), the coefficients are not significantly different from zero, and in the three remaining societies, the effect is very weak. From the findings presented in Table 2, we can conclude that both hypotheses H2a and H2b are falsified: in the eight West European societies under study, religious people are neither more nor less likely to adopt negative attitudes towards immigrants than their agnostic compatriots once the background variables are controlled for.

 

[TABLE 2 ABOUT HERE]

From Table 2 alone, one might be tempted to conclude that religiosity has no political consequences in Western Europe’s secularised societies. However, Table 3, which shows the probit regression of Christian Democratic / conservative party identification on religiosity as well as on the set of socio-demographic variables, indicates that this assertion would be incorrect: religiosity continues to have a huge impact on one’s likelihood of identifying with a Christian Democratic or conservative party even if the effects of socio-demographic variables are controlled for. The coefficients are substantial and significant in all countries, although it is interesting to note that the effect is a little weak in Italy and is unusually strong in the Netherlands. In the Netherlands the effect is substantially stronger for Catholics than it is for Protestants, while in Switzerland it is marginally stronger for Catholics than it is for Protestants (not shown as a table). Of course, with reference to our hypotheses, the strong impact of religiosity on party identification is a necessary but not a sufficient condition for the validity of Hypothesis H1, which suggested that religious people are less likely to vote for the radical right because they are firmly attached to Christian Democratic or conservative parties.

 

[TABLE 3 ABOUT HERE]

 

Table 3 also shows that men are more likely to identify with a Christian Democratic or conservative party than women. That said, since men are less religious in all countries, the direct positive effect of gender on party identification will often be effectively neutralised (in Belgium, France, and Norway) or even outweighed (in Denmark) by a negative indirect effect of gender via religiosity. The effect of class is negative throughout Western Europe, but is only significant in the two Scandinavian countries, where it most likely reflects the strength of the labour/capital cleavage. As for education, its effect is significantly positive in Austria, France, and Norway, but insignificant in all other countries. Finally, the effect of age is significant only in France, where it is huge. Again, this is after controlling for religiosity, which is already positively related to age, meaning that the direct and indirect effects of age will reinforce each other.

 

The (residual) correlation between identification with a Christian Democratic / conservative party and radical right-wing attitudes is negligible in all countries (see Table 4). This implies that supporters of these parties are neither more nor less likely to adopt negative attitudes towards immigrants than other voters once religiosity and socio-demographics are held constant.

 

[TABLE 4 ABOUT HERE]

 

Table 5 shows the probit regression of a vote for a party of the radical right on radical right-wing attitudes, religiosity, party identification, and the standard set of socio-demographic variables. A first observation is that the well-known effects of gender, age, class, and education are not significantly different from zero in most countries. The obvious explanation for this finding is that the strong effects of these socio-demographic attributes often found in studies of the radical right vote basically reflect the group differences in the strength of right wing attitudes that can be discerned from Table 2. That is, while education, for example, has a massive impact on attitudes, which in turn substantially affects the vote, the correlation between education and the vote disappears once attitudes are controlled for.

 

[TABLE 5 ABOUT HERE]

 

The explanatory power of attitudes is all the more evident in Table 5 if we look at the coefficients of radical right-wing attitudes. These are significant, large, and within the same range in seven of the eight countries. Table 5 therefore confirms that radical right-wing attitudes are a powerful predictor of the radical right vote, and that support for these parties should not be interpreted as a non-ideological, protest vote (van der Brug et al. 2000; van der Brug and Fennema 2003). The only exception here is Italy, where the effect is rather weak and is insignificant. This can be explained in part by the fact that the vast majority of Italian radical right-wing voters voted for the Alleanza Nazionale – a party has moderated its profile in recent years and that historically displayed limited hostility to foreigners in its ideology anyway (Carter 2005; Newell 2000).

 

The direct effect of religiosity on the probability of voting for a radical right party is less uniform across our countries. In Italy, religiosity has a borderline significant negative impact, while in Switzerland (where the effect is virtually identical for Catholics and Protestants) and France being religious clearly raises the probability of a radical right vote. Put differently, this indicates that in Switzerland and France the radical right appeals to religious voters net of them being encapsulated by Christian Democratic or conservative parties and of them being more or less anti-immigrant than other people. While there is no obvious explanation for this in the case of the Swiss SVP, the findings for France are in line with the FN’s appeals to a small but distinct fundamentalist Catholic constituency. In the five other countries, religiosity has no significant direct effect on the likelihood of voting for a radical right party – a finding that lends support to Hypothesis H3.

 

Finally, Table 5 indicates that the effects of identifying with a Christian Democratic or conservative party on the likelihood of voting for a party of the radical right are negative and often very large, although they are not significant in three of the eight countries under study. Combined with the results shown in Table 3, this provides further evidence for the validity of Hypothesis H1: in many cases, religious people are less likely to vote for the radical right because they are firmly attached to Christian Democratic or conservative parties.

 

[TABLE 6 ABOUT HERE]

 

From our model we can conclude that religiosity does play a significant role in explaining the radical right vote in Western Europe but that the picture is somewhat more complex than the (early) psychological research would suggest. In a bid to disentangle the various mechanisms, Table 6 illustrates the direct, indirect and total effects of religiosity on the likelihood of casting a vote for a party of the radical right in all eight countries under study. The first row of the table shows that the effect of religiosity via party identification is (often strongly) negative in all countries and significantly so in five of eight. By contrast, the second row illustrates that the effect of religiosity via radical right-wing attitudes is mostly weak and insignificant. The sum of these indirect effects (reported in the third row) is negative in all countries and significantly so in five of them. The direct effect of religiosity on the likelihood of casting a vote for a party of the radical right is reported in the fourth row of the table, which repeats the information from Table 5 above. The direct effect of religiosity is not uniform across the countries: in five of the eight societies it is not significant, whereas in France and Switzerland it raises the probability of a radical right vote, and in Italy it lowers this probability. These findings clearly highlight the importance of national contexts, and underline just how much religiosity, and indeed what it means to be religious, are shaped by distinct national influences. The final row of Table 6 reports the total effect of religiosity (indirect and direct). This is negative and significant in five countries, is negative and borderline-significant in Austria, and is not significantly different from zero in France and Switzerland.

 

 

Conclusion

 

The question that this article set out to investigate was whether religiosity influences the likelihood of an individual casting a vote for a party of the radical right in Western Europe. Our interest in this issue was guided by existing bodies of literature that led us to believe that a link between religiosity and radical right voting might well exist and by the fact that very few comparative studies have examined the subject. In an attempt to answer our question, we specified four separate hypotheses regarding the relationship between religiosity and voting for a radical right party. These enabled us to untangle the different effects that religiosity has on the radical right vote. In the first instance we suggested that religiosity might prevent people from voting for the radical right because religious people tend to develop an identification with a Christian Democratic or conservative party, and are thus simply not available to the parties of the radical right (Hypothesis H1). We also proposed that religiosity might have an effect on the support for the parties of the radical right via attitudes, and that this effect could either be negative (Hypothesis H2a) or positive (Hypothesis H2b). Lastly, we suggested that once attitudes and socio-demographic attributes are controlled for, there would be no substantial relationship between religiosity and the radical right vote (Hypothesis H3).

 

Somewhat surprisingly, this last hypothesis is not born out in practice in three of the eight countries, where there are significant direct effects of religiosity. There is no obvious explanation for the moderate negative direct effect of religiosity on the likelihood of a radical right vote in Italy, or its clearly stronger positive effect in Switzerland. By contrast, however, the positive effect of religiosity on the likelihood of a vote for the radical right in France is more easily accounted for. Not only has the Front National always taken a tough stand on issues such as abortion, homosexuality and the role of the church, but the party also has links with ultra-Catholic groups opposed to the church’s alleged ‘liberalism’ (Minkenberg 2003; Veugelers 2000). While studies of the Front National’s electorate demonstrate that most of its voters are overwhelmingly attracted by the party’s stance on immigration and are unconcerned about issues related to the church and its traditional teachings, and while the official church has become a leading critic of the FN’s anti-minority policies (Mayer and Perrineau 1992; Veugelers 2000), it is quite possible that these elements of the party’s appeal are attractive to a small segment of Catholic fundamentalists.

 

It also transpires that neither Hypothesis H2a nor Hypothesis H2b is born out in practice. We found no evidence that religious people are less likely to vote for the radical right because they are more altruistic, tolerant and compassionate and thus less likely to espouse negative attitudes towards immigrant; and nor did we find evidence to support the contrary suggestion that such people are more likely to vote for these parties because their religiosity is linked to higher levels of prejudice. While the second link in this causal chain (that anti-immigrant attitudes are very strong predictors of radical right voting) is confirmed in our findings (except in Italy, where, it has been argued, the AN is substantively different from other radical right parties), the first link is not: we found no relation between religiosity and anti-immigrant attitudes. All the effects were either statistically insignificant or irrelevant in substantial terms.

 

Of course, whether the absence of an overall relationship between religiosity and anti-immigrant sentiment is due to different mechanisms that counter-balance each other or to a true non-relationship cannot be ascertained with the data at hand. Yet, if we accept the absence of a link between religiosity and anti-immigrant attitudes at face value, this is clearly at odds with the findings of the earlier literature, and thus raises interesting questions. Setting aside concerns over the conceptual and methodological rigour of the early studies, one possible explanation for this contradiction would be that religiosity and ethnocentrism may well have been linked when these previous analyses were carried out (mainly in the 1950s and 1960s), but that this relationship has since waned and disappeared. Indeed, religious teachings, values and convictions are unlikely to have remained unaffected by social change, secularization and globalization, and it is thus very likely that belief systems are today less ‘closed’ than they used to be, and religious outlooks less ‘particularistic’. Yet the problem with this line of reasoning is that, everything else being equal, we would expect to have seen greater support for parties of the radical right in the 1950s and 1960s as compared to today. And this is clearly not the case: the radical right has been electorally more successful in the last two decades than at any point since World War Two.

 

Perhaps then the explanation is not temporal but geographical. Indeed, the vast majority of the studies that pointed to a link between religiosity and ethnocentrism were carried out in the US and it may well simply be that, while there was a relationship between religiosity and ethnocentrism among these respondents, that same relationship does not exist within West European electorates. This of course, once again, points to the importance of national contexts, both in terms of what religion means and entails in different societies and in terms of its manifestation and representation in the political system.

 

Clearly we can only speculate about the reasons why we found no link between religiosity and anti-immigrant attitudes and, as we noted above, it could be that there are different relationships between religiosity and anti-immigrant sentiment that actually counter-balance each other. From our more narrow perspective, however, regardless of this relationship, we can confidently conclude that in the societies under study, religiosity does not affect the vote for the radical right because of any influence religiosity might have on anti-immigrant attitudes.

 

Attitudes, however, remain crucial. Indeed, while the first link in our suggested causal chain (that religious people have either higher or lower levels of anti-immigrant sentiment) was falsified by our findings, the second was not. Like others (van der Brug et al. 2000; van der Brug and Fennema 2003), we found that negative attitudes towards immigrants are very strong predictors of radical right voting. Our analyses thus provide further evidence that voters who vote for parties of the radical right are doing so because they agree with the policies of these parties, and in particular with their anti-immigration appeals.

 

In contrast to H3, H2a and H2b, Hypothesis H1 is borne out in practice: in all countries religiosity has a substantial and statistically positive effect on the likelihood of a voter identifying with a Christian Democratic or conservative party. This in turn massively reduces the likelihood of casting a vote for a party of the radical right in many countries. We therefore conclude that ‘good Christians’ are neither especially tolerant towards ethnic minorities nor attracted by the radical right’s anti-immigrant rhetoric. Rather, to a large degree, they are simply still attached to Christian Democratic or conservative parties, and although they do not necessarily vote for these parties, this attachment ‘vaccinates’ them against voting for a party of the radical right (see Scarbrough 1984 on this idea of ‘vaccination’ in an electoral context).

 

This demonstrates that religiosity continues to be an important predictor of electoral choice. Yet, this ‘vaccine effect’ is likely to become weaker with time due to general de-alignment trends induced by social modernization and value change. Just as the parties of the mainstream left can no longer count on a traditional base of working class voters, Christian Democratic and conservative parties are today faced with fewer religious voters than they once were. Thus, in spite of still being able to ‘encapsulate’ religious voters, this natural reservoir of support is shrinking. All other things being equal, therefore, this points to an increase in the potential of radical right parties.

 

Acknowledgements

 

We would like to thank John Bartle, Thomas Poguntke, Elinor Scarbrough and Jack Veugelers for their valuable comments and suggestions on an earlier version of this article. We are also grateful to two anonymous reviewers and the editor of this journal for their helpful comments. Of course, the usual disclaimer applies.

 

 

Notes

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Tables

 

 

Table 1:        Determinants of religiosity

 

Religiosity on… Austria Belgium Denmark France Italy Neths. Norway Switz.
Gender -0.28* -0.38* -0.51* -0.37* -0.55* -0.23* -0.46* -0.30*
(0.06) (0.06) (0.07) (0.07) (0.07) (0.05) (0.06) (0.06)
Education -0.03 -0.00 -0.02 0.02 -0.07* -0.02 0.05 -0.09*
(0.02) (0.02) (0.04) (0.02) (0.04) (0.02) (0.03) (0.03)
Class 0.03 -0.02 0.10 -0.12 -0.11 0.04 0.07 0.05
(0.07) (0.07) (0.07) (0.08) (0.09) (0.06) (0.09) (0.07)
Age 30-65 0.58* 0.43* 0.52* 0.33* 0.32* 0.21* 0.37* 0.66*
(0.09) (0.09) (0.10) (0.09) (0.10) (0.09) (0.08) (0.12)
Age over 65 0.79* 1.31* 0.98* 1.08* 0.67* 0.65* 0.90* 1.00*
(0.11) (0.11) (0.12) (0.11) (0.12) (0.11) (0.09) (0.13)

Notes: Entries are unstandardized coefficients; standard errors are in brackets, *: p<.05

 

 

 

 

Table 2:        Determinants of radical right attitudes

 

Radical right attitudes on… Austria Belgium Denmark France Italy Neths. Norway Switz.
Religiosity 0.04 -0.02 -0.07* 0.06* 0.02 -0.03 0.07* 0.02
(0.03) (0.03) (0.03) (0.03) (0.04) (0.03) (0.03) (0.03)
Gender 0.05 -0.07 0.11 -0.06 0.03 -0.04 0.16* -0.06
(0.05) (0.06) (0.06) (0.07) (0.07) (0.05) (0.05) (0.06)
Education -0.30* -0.20* -0.34* -0.23* -0.23* -0.26* -0.33* -0.21*
(0.02) (0.02) (0.04) (0.02) (0.04) (0.02) (0.03) (0.03)
Class 0.27* 0.25* 0.15* 0.17* 0.31* 0.10 0.15* 0.26*
(0.07) (0.06) (0.07) (0.08) (0.10) (0.06) (0.06) (0.07)
Age 30-65 0.26* 0.19* 0.17 0.25* -0.23* -0.01 0.04 0.03
(0.08) (0.08) (0.09) (0.09) (0.10) (0.09) (0.07) (0.09)
Age over 65 0.60* 0.30* 0.56* 0.34* -0.32* 0.19 0.43* 0.30*
(0.11) (0.10) (0.11) (0.11) (0.13) (0.10) (0.09) (0.11)

Notes: Entries are unstandardized coefficients; standard errors are in brackets, *: p<.05

 

 

 

 

Table 3:        Determinants of Christian Democratic / conservative party identification

 

CD PID on… Austria Belgium Denmark France Italy Neths. Norway Switz.
Religiosity 0.53* 0.66* 0.38* 0.36* 0.27* 1.01* 0.48* 0.61*
(0.05) (0.07) (0.06) (0.06) (0.07) (0.07) (0.04) (0.09)
Gender 0.28* 0.30* 0.14 0.21* 0.28* 0.39* 0.29* 0.46*
(0.09) (0.10) (0.13) (0.10) (0.14) (0.09) (0.08) (0.14)
Education 0.10* -0.01 -0.00 0.09* 0.04 0.02 0.16* 0.10
(0.04) (0.04) (0.06) (0.03) (0.06) (0.04) (0.04) (0.07)
Class -0.03 -0.23 -0.47* 0.14 -0.28 -0.15 -0.25* -0.06
(0.10) (0.13) (0.14) (0.12) (0.15) (0.10) (0.13) (0.15)
Age 30-65 0.19 0.11 0.06 0.53* 0.18 0.07 0.02 -0.36
(0.14) (0.16) (0.21) (0.16) (0.17) (0.14) (0.11) (0.22)
Age over 65 0.15 0.33 0.29 0.80* 0.07 0.14 -0.01 -0.11
(0.17) (0.19) (0.23) (0.19) (0.23) (0.16) (0.14) (0.23)

Notes: Entries are unstandardized probit coefficients; standard errors are in brackets, *: p<.05

Table 4: Correlation of Christian Democratic / conservative party identification and radical right attitudes

 

Correlation with… Austria Belgium Denmark France Italy Neths. Norway Switz.
Rad right att 0.13* -0.08 0.13* 0.19* 0.13 -0.01 -0.04 -0.04
(0.04) (0.05) (0.06) (0.05) (0.07) (0.04) (0.04) (0.06)

Notes: Entries are correlations (Pearson); standard errors in brackets, *: p<.05.

 

 

 

 

Table 5:        Determinants of radical right voting

 

Radical right voting on… Austria Belgium Denmark France Italy Neths. Norway Switz.
Rad right att 0.72* 0.59* 0.60* 0.65* 0.19 0.62* 0.59* 0.50*
(0.24) (0.08) (0.07) (0.15) (0.11) (0.07) (0.07) (0.10)
Religiosity 0.28 0.03 -0.10 0.31* -0.20* 0.27 0.06 0.42*
(0.24) (0.18) (0.09) (0.13) (0.10) (0.14) (0.07) (0.19)
Gender 0.52 0.33 0.24 0.50* -0.11 0.23 0.39* 0.55*
(0.30) (0.18) (0.14) (0.21) (0.29) (0.13) (0.12) (0.22)
Education 0.19 -0.06 -0.13 0.03 0.05 0.00 -0.06 -0.03
(0.12) (0.07) (0.08) (0.07) (0.25) (0.04) (0.06) (0.07)
Class 0.00 0.03 0.01 0.35 0.19 -0.14 0.18 0.17
(0.25) (0.17) (0.16) (0.27) (0.55) (0.13) (0.15) (0.15)
Age 30-65 -0.26 -0.09 -0.22 0.72 0.15 -0.04 -0.30* -0.16
(0.28) (0.19) (0.17) (0.39) (0.46) (0.16) (0.15) (0.32)
Age over 65 -0.15 -0.45 -0.17 0.35 0.47 -0.28 -0.52* 0.07
(0.34) (0.31) (0.22) (0.40) (0.55) (0.19) (0.19) (0.34)
CD PID -0.92* -0.40 -0.17 -0.83* -0.26 -0.50* -0.61* -0.69*
(0.49) (0.25) (0.13) (0.22) (0.28) (0.12) (0.12) (0.22)

Notes: Entries are unstandardized probit coefficients; standard errors are in brackets, *: p<.05.
Test for CD PID is one-tailed.

 

 

 

 

Table 6:        Decomposition of the effect of religiosity

 

Religiosity on radical right voting Austria Belgium Denmark France Italy Neths. Norway Switz.
Via CD PID -0.48* -0.26 -0.06 -0.30* -0.07 -0.51* -0.29* -0.41*
(0.26) (0.17) (0.05) (0.09) (0.08) (0.13) (0.07) (0.16)
Via Rad right att 0.03 -0.01 -0.04* 0.04 0.00 -0.02 0.04* 0.01
(0.02) (0.02) (0.02) (0.02) (0.01) (0.02) (0.02) (0.02)
Total indirect -0.46 -0.27 -0.11* -0.26* -0.06 -0.53* -0.26* -0.40*
(0.25) (0.17) (0.05) (0.08) (0.08) (0.13) (0.07) (0.16)
Direct 0.28 0.03 -0.10 0.31* -0.20* 0.27 0.06 0.42*
(0.24) (0.18) (0.09) (0.13) (0.10) (0.14) (0.08) (0.19)
Total -0.18 -0.25* -0.21* 0.06 -0.26* -0.26* -0.19* 0.01
(0.10) (0.08) (0.08) (0.10) (0.09) (0.06) (0.05) (0.07)

Notes: Entries are unstandardized coefficients; standard errors are in brackets, *: p<.05.

Test for effect via CD PID is one-tailed.

 

 

Protest, Neo-Liberalism or Anti-Immigrant Sentiment: What Motivates the Voters of the Extreme Right in Western Europe?

 

1 The Research Problem: What Motivates the Voters of the Extreme Right?

The so-called “third wave” of post-war right-wing extremism (Beyme 1988) in Western Europe caught comparative political science by surprise. After the Second World War, the Extreme Right in Western Europe had been associated with the atrocities of the Nazis and their puppet regimes (Rydgren 2005) and was therefore politically isolated and insignificant in most countries of the region. But from the early 1980s on, parties that were dubbed as “Extremist”, “Radical”, “Populist” or “New” Right or any combination thereof1 and had been located at the margins of the political systems suddenly proved highly successful at the polls in countries such as Austria, Belgium, Denmark, France, Italy, Norway, and Sweden.

The diversity of these parties looked somewhat bewildering. Some had rather obvious connections with the old inter-war right while others qualified as “modern” (Ignazi 2002). Two of them pursued a separatist agenda (the Vlaams Blok and the Lega Nord) whereas the majority was firmly committed to national unity, and some of them had been founded as early as in the 1940s (the Italian MSI) whereas others (most notably New Democracy in Sweden) were only formed shortly before their first electoral successes.

But problems of terminology and idiosyncratic features notwithstanding, it soon became clear that these parties had some important commonalities and should be grouped into a single party family (Mudde 1996). While the members of this family may lack a common party label, and while there are few institutionalised structures to facilitate their transnational co-operation – two criteria that are frequently employed for party family membership, see Mair and Mudde 1998: 214-215 – the sociological profiles of their respective electorates turned out to be very similar. Moreover, the parties of the Extreme Right share a number of ideological features, in particular their concern about immigration, which became the single most important issue for these parties from the late 1980s on (Hainsworth 1992Kriesi 1999van der Brug and Fennema 2003).2

There is less agreement, however, as to what motivates the voters of the Extreme Right. In many of the earlier accounts, the notion of a (pure) “protest vote” features prominently. While there is no universally accepted definition of what constitutes a protest vote (but see Kang 2004), this literature suggests that protest reflects an unsatisfactory performance of the political system. Protest is therefore disconnected from ideology and should primarily be understood as “a vote against things” (Mayer and Perrineau 1992: 134). In a similar fashion, van der Brug and Fennema (2007: 478-479) argue that “the prime motive behind a protest vote is to show discontent with the political elite”, whereas political attitudes would be of less importance. This interpretation fits neatly into the discourse on anti-party sentiment that gained prominence in the early 1990s (Poguntke and Scarrow 1996), and there can be little doubt that at least some of the Extreme Right parties could benefit from widespread feelings of distrust and disaffection with the established parties.

The more recent literature, however, acknowledges that quite often, protest is not un-ideological at all but clearly directed “against the policy or the absence of policy in this respect [imigration and safety]” (Swyngedouw 2001: 218-219). Consequentially, the vast majority of comparative studies of the Extreme Right vote now adopt a theoretical framework that is based on the notion of a conflict between non-Western immigrants and the indigenous population over scarce resources (jobs, welfare benefits). Prominent examples of this approach include Jackman and Volpert (1996), Knigge (1998), Lubbers et al. (2002), Golder (2003a), and Arzheimer and Carter (2006), who all analyse the joint impact of immigration and unemployment on the electoral returns for the Extreme Right. More recently, Swank and Betz (2003) and Arzheimer (2008) have introduced the level of welfare benefits as an additional mediating variable.

Given the findings from this literature and the importance that the issue of immigration has gained for the parties of the Extreme Right, it makes obvious sense to assume that the voters of the Extreme Right are primarily motivated by concerns about immigration. Due to data restrictions, however, there is surprisingly little empirical evidence to support this view. The four studies by Jackman and Volpert, Knigge, Golder, and Swank and Betz are based on polity-level data alone and therefore have to take the anti-immigrant sentiment of the Extreme Right voters for granted. But even those comparative analyses that employ micro-data either assume a link between anti-immigrant sentiment and the Extreme Right vote or do have to rely on sub-optimal indicators.

Arzheimer and Carter (2006), for instance, present a hybrid model of Extreme Right voting that combines variables measured at the micro-level with information on the polity-level to capture the effects of “Political Opportunity Structures” on the individual vote. But this model does not include any items on individual political attitudes because the national election surveys on which their analysis is based “do not provide adequate data on attitudes” (Arzheimer and Carter 2006: 425). Rather, Arzheimer and Carter treat socio-demographic indicators like age, gender, and formal education as proxies for political preferences and values that might or might not dispose a respondent to vote for the Extreme Right (Arzheimer and Carter 2006: 421-422).

In a similar fashion, Lubbers et al. (2002: 357) estimate a complex multi-level model of Extreme Right voting. But because they use various data sources, they have to rely on single measure for anti-immigrant sentiment that is common to these data-sets, namely the question whether the respondents feels that “there are too many immigrants” in the country. While this is obviously a much more direct approach to the alleged link between anti-immigrant sentiment and the Extreme Right vote, operationalising a complex phenomenon like anti-immigrant sentiment with a single indicator is risky because this variable will be subject to both systematic and random measurement error. Likewise, even the useful study by van der Brug and Fennema (2003) that focuses exclusively on the question of whether the vote for the Extreme Right should be considered a “protest vote” relies on a single indicator to assess the impact of anti-immigrant sentiment on the Extreme Right vote, namely the subjective importance and satisfaction with immigration policies.

While anti-immigrant sentiment and (to a lesser degree) notions of “pure protest” dominate the recent discussion, two early but very influential accounts of the the “third wave” provided a rather different explanation for the success of the Extreme Right. Both Betz (1994) and Kitschelt (1995) claim that economic (neo-)liberalism is the key ingredient in the Extreme Right’s electoral “winning formula” (Kitschelt 1995: viii). According to them, “modern” parties like the Freedom Party in Austria or the Front National in France are enormously successful because they mix xenophobic statements with an attack on high taxation, the welfare state and its bureaucracy. Such a program would appeal to working class and lower middle class voters who feel that they do not benefit from “big government” but are likely to suffer from comparative disadvantages in a globalising labour market. More “traditional” Extreme Right parties like the German DVU and Republikaner, however, would never attract a similarly large constituency because they were wedded to the welfare policies of the inter- and postwar Extreme Right.

With the benefit of hindsight, the Extreme Right’s involvement with neo-liberal policies during the early 1990s now looks more like a brief fling. Consequentially, Betz (2003) has altoghether abandoned the idea that the Extreme Right does seriously pursue a “neo-liberal” agenda or has done so in the past, while Kitschelt has modified his original ideas considerably (McGann and Kitschelt 2005). Given the professional stature of both authors and the impact their respective monographs had on the field, an empirical test of the “winning formula” thesis is, however, overdue.

To summarise, while anti-immigrant sentiment has emerged as the most prominent motivation behind the Extreme Right vote in Western Europe, alternative accounts do exist and adequate tests of the respective causal links have by and large been restricted to a host of national studies (e.g. Billiet and Witte 1995Clark and Legge 1997Mughan and Paxton 2006). This is obviously problematic because each of these studies uses a different set of indicators, thereby rendering comparisons over time and countries invalid.

Fortunately, comparable data on attitudes towards immigrant as well as on electoral behaviour have recently become available with the first round of the European Social Survey (ESS). The aim of this article is therefore to make use of these data for modelling the effect of protest, immigrant sentiment and economic liberalism on the Extreme Right vote while at the same time controlling for a larger number of background variables than previous studies.

2 Data, Model and Methodology

 


Figure 1: A Simplified Model of the Extreme Right Vote in Western Europe

 


Data for the present study were collected in 2002/2003 under the auspices of the European Social Survey project. Of the 22 countries covered by this survey, seven were selected that have witnessed substantial support for the Extreme Right in recent years: Austria, Belgium, Denmark, France, Italy, the Netherlands, and Norway.3 While Golder (2003b) has argued that polity-level studies of the Extreme Right should also look at the “failed cases” (e.g. Spain or the UK) to avoid selection bias, it makes no sense to include them in micro-level analysis. If no one reports support for the Extreme Right, then there is simply nothing to model.4

Figure 1 represents the basic structure of the model. “Vote” is a dummy variable that takes the value of 1 if a respondent has voted for a party of the Extreme Right (see footnote 1) and the value of 0 if he or she has abstained or voted for another party. Vote is regressed on a a number of control variables as well as on the standard indicator (Arzheimer 2002) for non-ideological protest motives (“And on the whole, how satisfied are you with the way democracy works in this country?”), on sentiment towards immigrants, and on economic liberalism, thereby providing a direct test for the three most popular hypotheses about the motives or Extreme Right voters.

 


Culture How important should it be for immigrants to be committed to the way of life in [country]
Wages Average wages and salaries are generally brought down by people coming to live and work here
Skilled Labour People who come to live and work here help to fill jobs where there are shortages of workers
Jobs Would you say that people who come to live here generally take jobs away from workers in [country], or generally help to create new jobs?
Social Security Most people who come to live here work and pay taxes. They also use health and welfare services. On balance, do you think people who come here take out more than they put in or put in more than they take out?
Economy Would you say it is generally bad or good for [country]’s economy that people come to live here from other countries?
Cultural Threat Would you say that [country]’s cultural life is generally undermined or enriched by people coming to live here from other countries?
Quality of Life Is [country] made a worse or a better place to live by people coming to live here from other countries?
Crime Are [country]’s crime problems made worse or better by people coming to live here from other countries?
Labour Migration All countries benefit if people can move to countries where their skills are most needed
Multi-Culturalism It is better for a country if almost everyone shares the same customs and traditions
Religious Diversity It is better for a country if there are a variety of different religions
Linguistic Diversity It is better for a country if almost everyone is able to speak at least one common language
Immigration If a country wants to reduce tensions it should stop immigration
Fair Share [Country] has more than its fair share of people applying for refugee status

Scales for “Wages”, “Skilled Labour”, “Labour Migration”, “Multi-Culturalism”, “Religious Diversity”, “Linguistic Diversity”, “Immigration” and “Fair Share’ run from 1 (“Agree Strongly”) to 5 (“Disagree Strongly”). All other scales run from 1 to 11.

Where necessary, scales were reversed so that high values refer to the pro-immigrant position.

Table 1: Indicators for Immigrant Sentiment


In the literature, immigrant sentiment is often portrayed as a complex phenomenon (Mughan and Paxton 2006). Moreover, given the different levels and patterns of immigration in Wester Europe, one cannot take for granted that interviewees from different countries respond to any given indicator in exactly the same way. Therefore, immigrant sentiment is conceptualised as a “latent” (not directly observable) variable in the model.5 Having a separate measurement model for this attitude makes the overall model more robust and allows one to assess the reliability of the indicators in comparative perspective.

The 15 indicators selected for the measurement of anti-immigrant attitudes (see Table 1) reflect two major components of anti-immigrant sentiment, namely perceptions of material and cultural threats. However, while these two sub-dimensions are conceptually separable (Mughan and Paxton 2006: 342-343), the respective items display a very high degree of intercorrelation in all countries under study and are therefore interpreted as indicators for a single latent variable. To ease the interpretation, items were rescaled so that positive values of the latent variable correspondent to pro-immigrant sentiment whereas negative values stand for anti-immigrant sentiment.

 


Income Equalisation The government should take measures to reduce differences in income levels
Government Intervention The less that government intervenes in the economy, the better it is for [country]
Trade Unions Employees need strong trade unions to protect their working conditions and wages

Scales run from 1 (“Agree Strongly”) to 5 (“Disagree Strongly”)

For “Government Intervention”, the scale was reversed so that high values refer to the economically liberal position, too.

Table 2: Indicators for Economic Liberalism


To test Betz’s and Kitschelt’s early hypothesis about the importance of pro-market attitudes for the Extreme Right vote, the model contains a second latent variable dubbed “Economic Liberalism”. It is constructed from three indicators that capture resistance against equalisation of incomes, against trade unions and against state intervention in the economy (see Table 2).

As outlined above, socio-demographic variables often play an important role as proxy variables for attitudes in the existing research on the Extreme Right because theory suggests various causal links between both groups of variables. For instance, ethnic competition theory suggests that higher levels of formal education should be associated with lower levels of anti-immigrant sentiment (because most non-Western migrants are unskilled) and therefore with a lower propensity to vote for the Extreme Right. Moreover, there is ample evidence that formal education promotes liberal values (e.g. Weakliem 2002), whose adoption should also reduce levels of anti-immigrant sentiment. Either way, once anti-immigrant sentiment is controlled for, socio-demographic variables should have only minimal direct effects on the vote.

Again, most existing research simply assumes that socio-demographics can be used as a proxy variables for anti-immigrant sentiment, precisely because good indicators for attitudes are not generally available. To test this assertion as well as to control for residual effects, the model contains a large selection of socio-demographic variables (gender, age, union membership, church attendance, class6, employment status, education7, household size, and relationship status8) that have been shown to have an effect on the Extreme Right vote in previous research. Both direct and indirect (via anti-immigrant sentiment and economic liberalism) effects link these variables and the vote.

Finally, it has been noted that the literature on the voters of the Extreme Right is empirically and analytically not well connected to the very large body of research on mainstream electoral behaviour (Arzheimer 2008). In a bid to overcome this unfortunate divide, two standard attitudinal measures were included in the model as additional controls: according to the Michigan school, party identification9 is the single most important predictor of electoral behaviour, whereas ideology (left-right self placement) plays a prominent role in spatial approaches to electoral behaviour that build on the work of Hotelling (1929) and Downs (1957). Not controlling for these important predictors could lead to significant bias in the results.

The presence of latent variables and the (block-causal)10 structure of the model call for Structural Equation Modelling (Kaplan 2000), a statistical technique that allows one to estimate the parameters for the relationships between several variables simultaneously. Estimation was carried out on a per-country basis so that the sign and strength of effects can be compared across polities.

While Structural Equation Modelling (SEM) is now a well-established technique, three complications remain. First, the vote for the extreme right is a dichotomous variable whereas SEM was originally developed for continuous variables. However, modern software allows one to specify a nonlinear link between a dichotomous variable and its antecedents to deal with this problem.11

Second, while levels of item non-response are generally very low in the ESS, even a small proportion of missing values adds up in a model with so many variables. To avoid bias, over-optimistic standard errors and the massive reduction of the sample size that would result from listwise deletion (i.e. complete case analysis), Multiple Imputation by Chained Equations (MICE, see van Buuren and Oudshoorn 1999) was applied to fill the gaps in the data with a range of plausible values. For each country, 21 separate imputations of the original data were created using Royston’s (2005) implementation of MICE in Stata. Since MICE is a stochastic procedure, the differences between these imputations reflect the uncertainty about the missing values. Results from the 21 separate analyses were then combined in Mplus according to the rules outlined in Rubin (1987). This somewhat complex procedure yields approximately unbiased parameter estimates and conservative standard errors that take the amount of missing data into account, thereby providing an additional margin of safety.

Finally, an (arbitrary) scale for the two latent variables (immigrant sentiment and economic liberalism) must be established to identify the model. This was done by assuming that these variables are standardised, i.e. that they have a mean of zero and unit variance.

3 Findings

3.1 Overall Fit and Measurement Models

 


AT BE DK FR IT NL NO
RMSEA 0.059 0.062 0.059 0.065 0.060 0.066 0.059
N 2080 1676 1404 1418 1155 2246 1928

Root Mean Squared Errors of Approximation (RMSEA) and number of observations (N), averaged over 21 imputations

Table 3: Fit of the Model


Amongst the many fit indices for Structural Equation Models that have been proposed in the literature, the Root Mean Squared Error of Approximation (RMSEA) is arguably the most popular at the moment because it has a well-known distribution and is less sensitive to the size of the sample than other measures (Garson 2008). Table 3 shows that in all seven countries, the RMSEA is well below the conventional threshold of 0.1 and actually comes very close to the value of 0.05, which indicates a very good fit.

Estimates for the measurement model for immigrant sentiment are equally encouraging (Table 4). All coefficients are statistically significant12 and have the correct sign. Moreover, for most indicators the parameters are roughly within the same range, implying that the respective indicators are more or less equivalent. Since the entries in Table 4 are really just unstandardised regression coefficients, their interpretation is straightforward. For instance, the last of the 15 items asks whether the respondent agrees that the country gets a “fair share” of refugees. If one now compares two respondents with average (0) and rather positive (1) feelings towards immigrants, this difference of one standard deviation on the latent variable results in a substantively higher (about 0.5 points on a five-point rating scale) level of agreement with the pro-refugee statement, with the strongest effect (0.558 points) in Austria and the weakest (0.316 points) in Italy.

 


Variable AT BE DK FR
Culture −1.095 (0.060) −0.655 (0.048) −1.063 (0.072) −0.990 (0.057)
Wages 0.385 (0.027) 0.438 (0.031) 0.246 (0.028) 0.487 (0.033)
Skilled Labour −0.173 (0.023) −0.232 (0.025) −0.207 (0.025) 0.023 (0.026)
Jobs 1.149 (0.039) 1.112 (0.042) 0.803 (0.040) 1.199 (0.051)
Social Security 1.489 (0.048) 1.096 (0.046) 1.084 (0.052) 1.291 (0.051)
Economy 1.434 (0.047) 1.382 (0.045) 1.635 (0.060) 1.568 (0.049)
Cultural Threat 1.618 (0.049) 1.156 (0.048) 1.566 (0.055) 1.723 (0.058)
Quality of Life 1.455 (0.039) 1.265 (0.040) 1.469 (0.046) 1.537 (0.043)
Crime 1.225 (0.041) 1.029 (0.044) 1.062 (0.050) 1.139 (0.053)
Labour Migration −0.196 (0.025) −0.191 (0.023) −0.112 (0.027) −0.079 (0.021)
Multi-Culturalism 0.603 (0.028) 0.478 (0.028) 0.578 (0.033) 0.506 (0.034)
Religious Diversity −0.507 (0.026) −0.302 (0.026) −0.466 (0.030) −0.383 (0.027)
Linguistic Diversity 0.243 (0.017) 0.071 (0.017) 0.107 (0.015) 0.107 (0.014)
Immigration 0.621 (0.028) 0.587 (0.026) 0.535 (0.033) 0.768 (0.036)
Fair Share 0.558 (0.026) 0.541 (0.024) 0.455 (0.031) 0.465 (0.027)
Variable IT NL NO
Culture −0.334 (0.067) −0.691 (0.038) −1.116 (0.054)
Wages 0.468 (0.037) 0.334 (0.021) 0.176 (0.018)
Skilled Labour −0.352 (0.028) −0.214 (0.020) −0.125 (0.017)
Jobs 0.767 (0.071) 0.667 (0.032) 0.719 (0.034)
Social Security 0.998 (0.061) 1.108 (0.041) 0.991 (0.041)
Economy 1.456 (0.062) 1.262 (0.036) 1.223 (0.038)
Cultural Threat 1.404 (0.071) 1.069 (0.037) 1.328 (0.043)
Quality of Life 1.199 (0.062) 1.141 (0.037) 1.157 (0.034)
Crime 0.863 (0.066) 1.000 (0.035) 0.794 (0.032)
Labour Migration −0.183 (0.026) −0.088 (0.019) −0.101 (0.019)
Multi-Culturalism 0.444 (0.033) 0.482 (0.024) 0.502 (0.025)
Religious Diversity −0.310 (0.031) −0.214 (0.019) −0.369 (0.019)
Linguistic Diversity 0.066 (0.021) 0.139 (0.013) 0.108 (0.012)
Immigration 0.649 (0.035) 0.541 (0.023) 0.454 (0.021)
Fair Share 0.316 (0.035) 0.474 (0.019) 0.388 (0.019)

Entries are unstandardised regression coefficients (WLSMV) based on 21 imputations. Standard errors in parentheses. See Table 1 in the appendix for the full text of the items

Table 4: Immigrant Sentiment: Measurement Model


The measurement model for economic liberalism (Table 5), however, works less well. More specifically, the item on government interventions is only loosely connected to the latent variable13 whereas the two other items perform well. This is probably due to an extremely skewed distribution of the responses: in all countries but Austria, majorities in excess of 70 per cent either support government interventions or are at least indifferent. However, since scepticism about government interventions obviously reflects the theoretical content of economic liberalism, the item was retained.

 


Variable AT BE DK FR
Income Equalisation 0.470 (0.054) 0.288 (0.037) 0.277 (0.072) 0.430 (0.053)
Government Intervention 0.161 (0.036) −0.187 (0.040) 0.036 (0.069) −0.086 (0.043)
Trade Unions 0.748 (0.082) 0.388 (0.050) 0.648 (0.206) 0.425 (0.051)
Variable IT NL NO
Income Equalisation 0.333 (0.066) 0.354 (0.036) 0.361 (0.031)
Government Intervention −0.033 (0.039) 0.015 (0.023) 0.010 (0.026)
Trade Unions 0.747 (0.146) 0.573 (0.056) 0.425 (0.036)

Entries are unstandardised regression coefficients (WLSMV) based on 21 imputations. Standard errors in parentheses.

Table 5: Economic Liberalism: Measurement Model


3.2 Antecedents of Economic Liberalism and Immigrant Sentiment

Table 6 presents the coefficients for the regression of economic liberalism on a range of socio-demographic control variables. Rather unsurprisingly, the unemployed, trade unionists, and members of the working class show substantively lower levels of economic liberalism than other respondents. Conversely, interviewees holding university degrees show disproportionate support for market capitalism. However, the relative and absolute strength of these effects varies considerably. In France, for instance, education is the key factor, whereas class and trade union membership are dominant in Denmark or Austria. Moreover, gender makes a significant difference in all countries but Italy: even if a whole host of other socio-demographics is controlled for, men tend to support market mechanisms more strongly than women.

 


Variable AT BE DK FR
Male 0.198 (0.064) 0.263 (0.109) 0.116 (0.110) 0.271 (0.101)
Age: 18-29 −0.157 (0.113) −0.103 (0.181) −0.292 (0.174) 0.061 (0.159)
Age: 30-45 0.041 (0.082) 0.403 (0.141) 0.244 (0.119) 0.062 (0.138)
Age: over 65 −0.012 (0.112) −0.795 (0.232) −0.123 (0.195) −0.019 (0.183)
Religion: none −0.005 (0.081) −0.082 (0.128) 0.061 (0.108) −0.093 (0.119)
Church Attendance 0.038 (0.023) −0.076 (0.047) 0.066 (0.050) 0.038 (0.043)
Trade Union Member −0.348 (0.085) −0.551 (0.132) −0.423 (0.143) −0.068 (0.220)
Petty Bourgeoisie 0.385 (0.158) 0.728 (0.234) 0.388 (0.224) 0.324 (0.232)
Working Class −0.427 (0.103) −0.822 (0.173) −0.485 (0.157) −0.042 (0.175)
Pensioner −0.101 (0.106) −0.086 (0.222) −0.287 (0.184) −0.192 (0.186)
Unemployed 0.075 (0.174) −1.132 (0.303) −0.098 (0.223) −0.173 (0.233)
University Degree 0.103 (0.093) 0.990 (0.170) 0.261 (0.140) 0.907 (0.131)
Household Size = 1 0.056 (0.109) −0.108 (0.202) 0.222 (0.206) 0.136 (0.194)
-Single −0.034 (0.102) −0.088 (0.160) 0.237 (0.199) 0.155 (0.180)
Variable IT NL NO
Male 0.113 (0.087) 0.326 (0.070) 0.427 (0.080)
Age: 18-29 −0.249 (0.153) −0.077 (0.134) 0.038 (0.127)
Age: 30-45 −0.280 (0.110) 0.102 (0.078) 0.107 (0.092)
Age: over 65 −0.135 (0.165) −0.219 (0.139) −0.332 (0.235)
Religion: none 0.067 (0.109) −0.002 (0.083) −0.062 (0.082)
Church Attendance 0.012 (0.028) −0.037 (0.026) 0.004 (0.034)
Trade Union Member −0.229 (0.119) −0.580 (0.091) −0.645 (0.086)
Petty Bourgeoisie 0.016 (0.134) 0.282 (0.235) −0.041 (0.216)
Working Class −0.464 (0.142) −0.245 (0.104) −0.160 (0.113)
Pensioner −0.343 (0.160) −0.003 (0.139) −0.090 (0.232)
Unemployed −0.246 (0.176) −0.194 (0.245) −0.409 (0.202)
University Degree 0.228 (0.146) 0.551 (0.083) 0.604 (0.094)
Household Size = 1 0.024 (0.169) −0.042 (0.148) 0.019 (0.157)
-Single −0.043 (0.109) −0.080 (0.137) 0.141 (0.139)

Entries are unstandardised regression coefficients (WLSMV) based on 21 imputations. Standard errors in parentheses.

Table 6: Regression of Economic Liberalism on Socio-Demographics


As discussed in sections 1 and 2, most comparative analyses of the extreme right vote in Western Europe rely on a putative link between socio-demographic indicators of group-membership on the one hand and anti-immigrant sentiment on the other. Table 7 demonstrates that this practice is justified, at least up to a degree: formal education emerges clearly as the single most important predictor of sentiment towards immigrants. In all seven countries studied here, respondents with a university degree report much more positive feelings towards immigrants than other interviewees. The difference is roughly equivalent to half a standard deviation of the latent variable and varies from 0.48 points in the Netherlands to 0.69 points in Austria. Moreover, even though education is controlled for, class has an effect, too: in most countries, members of the working class and the “petty bourgeoisie” display a much more negative attitude towards immigrants than other respondents. Other variables (unemployment in particular) have smaller and more erratic effects.

 


Variable AT BE DK FR
Male −0.010 (0.049) 0.076 (0.056) −0.102 (0.062) 0.097 (0.060)
Age: 18-29 0.318 (0.087) 0.193 (0.095) 0.205 (0.107) 0.513 (0.098)
Age: 30-45 0.229 (0.065) 0.137 (0.072) 0.057 (0.076) 0.417 (0.083)
Age: over 65 −0.301 (0.086) −0.160 (0.107) −0.148 (0.107) 0.044 (0.109)
Religion: none 0.280 (0.064) −0.020 (0.068) 0.102 (0.066) 0.019 (0.073)
Church Attendance 0.057 (0.018) 0.039 (0.022) 0.085 (0.029) 0.041 (0.025)
Trade Union Member −0.017 (0.060) −0.031 (0.064) −0.025 (0.076) 0.268 (0.114)
Petty Bourgeoisie −0.306 (0.118) −0.318 (0.123) −0.143 (0.153) 0.031 (0.164)
Working Class −0.454 (0.077) −0.314 (0.082) −0.206 (0.083) −0.293 (0.099)
Pensioner −0.254 (0.082) −0.338 (0.109) −0.337 (0.110) −0.137 (0.115)
Unemployed −0.243 (0.150) −0.400 (0.120) −0.176 (0.148) −0.156 (0.129)
University Degree 0.694 (0.072) 0.593 (0.090) 0.645 (0.086) 0.669 (0.077)
Household Size = 1 −0.157 (0.085) −0.222 (0.105) −0.052 (0.128) −0.116 (0.106)
-Single −0.082 (0.079) −0.171 (0.083) −0.051 (0.117) −0.166 (0.097)
Variable IT NL NO
Male −0.049 (0.071) 0.017 (0.050) −0.075 (0.053)
Age: 18-29 −0.046 (0.130) 0.195 (0.091) 0.135 (0.086)
Age: 30-45 0.154 (0.087) 0.242 (0.059) 0.107 (0.066)
Age: over 65 −0.166 (0.129) −0.315 (0.095) −0.346 (0.134)
Religion: none 0.300 (0.087) 0.078 (0.067) 0.117 (0.056)
Church Attendance 0.014 (0.023) 0.033 (0.021) −0.039 (0.023)
Trade Union Member −0.083 (0.100) 0.154 (0.060) 0.063 (0.054)
Petty Bourgeoisie −0.192 (0.107) −0.151 (0.122) −0.353 (0.166)
Working Class −0.444 (0.106) −0.221 (0.082) −0.256 (0.074)
Pensioner −0.115 (0.134) 0.062 (0.095) −0.340 (0.144)
Unemployed −0.383 (0.132) 0.253 (0.195) −0.197 (0.138)
University Degree 0.487 (0.152) 0.476 (0.060) 0.679 (0.064)
Household Size = 1 −0.092 (0.141) 0.178 (0.102) −0.045 (0.104)
-Single −0.059 (0.099) 0.084 (0.093) −0.087 (0.093)

Entries are unstandardised regression coefficients (WLSMV) based on 21 imputations. Standard errors in parentheses.

Table 7: Regression of Immigrant Sentiment on Socio-Demographics


3.3 The Extreme Right Vote in Comparative Perspective

Finally, Tables 8 and 9 presents the (probit) regression of the Extreme Right vote on socio-demographic and attitudinal variables. A first important finding is that once attitudinal variables are controlled for, in all of the seven countries socio-demographic variables have no significant effect on the vote whatsoever. Put differently, the large and persistent differences as regards the propensity of various social groups to vote for the Extreme Right that have been observed in national and comparative studies are entirely due to differences between these groups with respect to the five attitudinal variables included in the model.

As regards the protest motives, the effect of satisfaction with the way democracy works in one’s country is statistically insignificant in most countries. Only in Belgium and the Netherlands, there is a link between (dis)satisfaction and the Extreme Right vote. While the absolute value of the coefficient is small (-0.14 and -0.12 respectively), its potential impact is large because satisfaction was measured on a ten-point rating scale. However, the interquartile range for satisfaction in Belgium and the Netherlands (loosely speaking the difference between those who are fairly dissatisfied and those who are fairly satisfied) amounts to only 3 and 2 points respectively, which would result in a rather moderate impact on the likelihood of an Extreme Right vote. On balance, these results suggest that the role of “pure protest” motives is very limited.

Similarly, economic liberalism is obviously not a key ingredient in the electoral winning formula for the Extreme Right: its effects are insignificant in all countries. Crucially, the effect is negative (though statistically insignificant) for the voters of the French Front National, Kitschelt’s 1995 “master case” of a “new” rightist party.

On the other hand, positive sentiment towards immigrants generally exerts a significant negative effect on the vote. Put differently, concerns about immigrants and immigration policies emerge as major motivation for the voters of the Extreme Right in six out of seven countries. The single exception is Italy, where the effect is not significantly different from zero. This specific finding sheds an interesting light on the Alleanza Nazionale, whose supporters make up the vast majority14 of the Italian Extreme Right voters in the data set: first, even the Alleanza’s neo-fascist predecessor party MSI displayed only very limited hostility to foreigners (Newell 2000), and second, the party has moderated its profile so much in recent years that some scholars do not longer consider it as part of the Extreme Right. While one can obviously not judge a party by its voters, the results demonstrate that the Alleanza’s supporters are different in so far as they are apparently not particularly attracted by anti-immigrant rhetoric and policies. Rather, they seem to be motivated by their general left-right preferences and their identification with the party.

As regards ideology, the findings are similarly clear-cut: more right-leaning respondents are far more likely to vote for the extreme right even after immigrant sentiment is controlled for in all countries but Denmark and France, where the effect does not pass the conventional threshold of statistical significance. Again, this speaks against the idea that the voters of the Extreme Right are motivated by pure protest motives which are unrelated to policy considerations.

Finally, party identifications have a very strong and highly plausible effect on the Extreme Right vote: respondents who identify with these parties display a very high propensity to vote for them, whereas an identification with any other party acts as an effective deterrent. While this may seem fairly obvious (if not tautological), almost all existing analyses neglects the role of party identification. This is problematic precisely because party identification has such a strong effect on the vote. If this force is ignored, severe bias can result. Also, like with other parties, the match between party identification and voting behaviour is by no means perfect. The share of identifiers amongst the voters of the Extreme Right varies between 25 (Netherlands) and 67 (Italy) per cent, whereas between 54 (France) and 85 (Italy) per cent of the identifiers vote for the respective party.

This important role of party identification provides additional evidence against the pure protest hypothesis. Moreover, only when this strong yet imperfect link is controlled for, one can truly appreciate the importance the influence of immigrant sentiment and ideology: although the single most important predictor of the Extreme Right vote is statistically held constant, policy-related attitudes still exert a very strong influence.

 


Variable AT BE DK FR
Immigrant Sentiment −0.196 (0.059) −0.189 (0.057) −0.363 (0.051) −0.201 (0.063)
Economic Liberalism 0.144 (0.080) −0.025 (0.124) −0.046 (0.107) −0.147 (0.109)
PID: Extreme Right 2.251 (0.215) 1.820 (0.216) 1.928 (0.244) 1.410 (0.277)
PID: other −0.761 (0.212) −0.841 (0.287) −0.646 (0.153) −0.642 (0.218)
Left-Right Placement 0.157 (0.052) 0.110 (0.042) 0.068 (0.037) 0.088 (0.052)
Satisfied: democracy −0.050 (0.033) −0.139 (0.039) −0.030 (0.036) −0.075 (0.045)
Male 0.132 (0.156) 0.100 (0.175) 0.188 (0.141) 0.217 (0.169)
Age: 18-29 −0.065 (0.341) −0.033 (0.251) 0.127 (0.216) −0.336 (0.410)
Age: 30-45 −0.050 (0.220) 0.013 (0.210) −0.082 (0.194) 0.090 (0.217)
Age: over 65 −0.018 (0.233) −0.249 (0.338) 0.303 (0.242) 0.404 (0.358)
Religion: none 0.411 (0.199) 0.121 (0.179) 0.137 (0.152) −0.173 (0.224)
Church Attendance 0.080 (0.060) −0.065 (0.088) −0.117 (0.084) −0.015 (0.089)
Trade Union Member −0.044 (0.205) 0.039 (0.185) 0.221 (0.196) 0.250 (0.328)
Petty Bourgeoisie −0.080 (0.389) −0.001 (0.401) −0.153 (0.357) −0.381 (0.527)
Working Class 0.424 (0.225) 0.056 (0.218) 0.253 (0.196) −0.035 (0.244)
Pensioner 0.111 (0.254) −0.028 (0.308) 0.090 (0.252) −0.498 (0.340)
Unemployed 0.203 (0.514) 0.322 (0.321) 0.467 (0.299) −0.455 (0.495)
University Degree 0.238 (0.288) −0.329 (0.445) −0.727 (0.416) −0.037 (0.297)
Household Size = 1 0.108 (0.337) 0.347 (0.276) −0.322 (0.266) −0.052 (0.404)
-Single 0.173 (0.325) 0.139 (0.239) −0.238 (0.248) 0.072 (0.412)
Constant −3.052 (0.574) −1.742 (0.493) −1.469 (0.492) −1.581 (0.650)

Entries are unstandardised Probit regression coefficients (WLSMV) based on 21 imputations. Standard errors in parentheses.

Table 8: Regression of the Extreme Right Vote on Socio-Demographics and Attitudes I


 


Variable IT NL NO
Immigrant Sentiment −0.035 (0.066) −0.254 (0.041) −0.259 (0.044)
Economic Liberalism 0.148 (0.075) 0.028 (0.048) 0.087 (0.070)
PID: Extreme Right 2.498 (0.302) 1.434 (0.159) 1.363 (0.113)
PID: other −1.171 (0.467) −0.577 (0.091) −1.052 (0.150)
Left-Right Placement 0.173 (0.070) 0.150 (0.022) 0.152 (0.031)
Satisfied: democracy −0.023 (0.054) −0.117 (0.023) −0.018 (0.025)
Male 0.172 (0.208) 0.097 (0.087) 0.041 (0.104)
Age: 18-29 0.003 (0.511) −0.240 (0.175) −0.233 (0.148)
Age: 30-45 0.012 (0.324) −0.060 (0.101) −0.208 (0.130)
Age: over 65 0.288 (0.433) −0.202 (0.197) −0.294 (0.249)
Religion: none 0.240 (0.334) 0.156 (0.105) −0.079 (0.101)
Church Attendance −0.066 (0.087) −0.050 (0.039) −0.114 (0.041)
Trade Union Member −0.028 (0.514) −0.214 (0.116) 0.087 (0.110)
Petty Bourgeoisie 0.231 (0.336) 0.007 (0.150) −0.215 (0.252)
Working Class −0.300 (0.440) 0.081 (0.127) −0.014 (0.130)
Pensioner −0.244 (0.498) −0.001 (0.195) 0.498 (0.241)
Unemployed −0.383 (0.590) 0.419 (0.321) 0.265 (0.263)
University Degree 0.161 (0.431) −0.001 (0.121) −0.226 (0.160)
Household Size = 1 0.046 (0.523) 0.109 (0.184) −0.008 (0.178)
-Single 0.158 (0.410) 0.201 (0.169) 0.029 (0.152)
Constant −2.574 (0.667) −1.256 (0.273) −1.699 (0.317)

Entries are unstandardised Probit regression coefficients (WLSMV) based on 21 imputations. Standard errors in parentheses.

Table 9: Regression of the Extreme Right Vote on Socio-Demographics and Attitudes II



PIC

Figure 2: The Effect of Ideology, Immigrant Sentiment, and Party Identification on the Extreme Right Vote (No Party ID)



PIC

Figure 3: The Effect of Ideology, Immigrant Sentiment, and Party Identification on the Extreme Right Vote (Extreme Right Party ID)



PIC

Figure 4: The Effect of Ideology, Immigrant Sentiment, and Party Identification on the Extreme Right Vote (Other Party ID)


This is most readily seen when the findings are converted from the probit scale back to the “quantity of interest” (King et al. 2000), i.e. the probability of a vote for the Extreme Right. For instance, for a right-leaning (say 7 on the ideology scale) respondent from Norway who identifies with the Freedom Party (1) and has rather negative (-1) attitudes towards immigrants, one would simply multiply these values with their respective coefficients, add the constant and plug the result (-1.699 + 7 × 0.152 + 1 × 1.363 + -1 × -0.259 = 0.987) into the standard cumulative density function Φ to obtain the probability of an Extreme Right vote (0.838).15 Figures 24 show how this probability varies with ideology (the solid, short-dashed and long dashed lines), immigrant sentiment, and party identification.16

From Figures 2, it is readily apparent that both ideology and immigrant sentiment have a sizeable impact amongst non-partisans: for a right-leaning voter who dislikes immigrants, the probability of a vote for the Freedom Party quickly approaches 40 per cent, while this probability is 20 per cent or less for left-leaning voters, especially if they are favourably disposed towards immigrants.17 But even amongst those respondents who identify with the Freedom Party (see Figure 3), the probability of an Extreme Right vote is clearly less than 100 per cent and varies considerably with ideology and immigrant sentiment (Figure 3).18 Perhaps the most interesting constellation is depicted in Figure 4. Here, one can see that even respondents who identify with another party have a sizeable probability of voting for the Freedom Party, provided that they are right-leaning and strongly oppose immigration (cf. the upper-left corner of the graph). While such a vote would still be a rather rare event, the probability of an Extreme Right vote is considerably (i.e. roughly ten times) higher in this group than amongst those respondents who have a more favourable attitude towards immigrants (cf. the lower-right corner of the graph).

 


AT BE DK FR IT NL NO
Ratio 2.78 1.88 3.27 1.73 1.16 1.79 2.15

Entries are ratios of the expected vote shares amongst anti-immigrant (-1) and pro-immigrant (+1) centrist (5) citizens with no party identification. The (mostly insignificant) effects of all other variables were set to zero.

Table 10: The impact of immigrant sentiment amongst independents in comparative perspective


Further graphical comparisons between countries are hampered by the fact that the base level of Extreme Right support (as reflected by the constant in Tables 8 and 9) varies considerably, resulting in essentially flat lines for countries with low levels of Extreme Right support. Therefore, ratios of predicted vote shares were calculated to put the impact of immigrant sentiment into comparative perspective. These calculations focus on a group that is of particular interest for political strategists in all West European countries: centrist citizens who are not attached to a particular party. In Norway, for instance, members of this group who display a rather positive (+1)19 attitude towards immigrants have a probability of roughly 12 per cent to vote for the Freedom Party. But for members of the same group who clearly dislike immigrants (-1 on the scale), the probability of a Freedom Party more than doubles. As can be gleaned from Table 10, for most countries this ratio is roughly in the same range.

The obvious exception is Italy, were immigrant sentiment makes virtually no difference as regards the electoral prospects of the Extreme Right.20 On the other hand, in Austria and Denmark support for the Extreme Right roughly triples with increasing anti-immigrant sentiment. While this figure might be slightly misleading in the case of Austria because support for the Freedom Party was generally very low amongst centrist independents so that the political impact of immigrant sentiment must remain limited within this group, attitudes vis-a-vis immigrants make all the difference in Denmark. Here, even those independent centrists who hold favourable views of immigrants have a seven per cent probability of voting for the Extreme Right. Consequentially, the model predicts that about one in five independents who strongly dislike immigrants but have otherwise centrist political preferences will vote for the Extreme Right.

4 Conclusion

Parties of the “Extreme”, “Radical” or ‘Populist” Right have become a permanent feature of West European politics, and since the mid-1980s, immigration has been the most important issue for them. Recent research has linked the levels of support these parties receive to polity-level variables such us unemployment and immigration. However, comparative micro-level evidence on the motives of their voters is still scarce.

Using recent survey data and a more appropriate measurement model than previous research, this article has demonstrated that Kitschelt’s 1995 hypothesis about the importance of neo-liberal policy preferences is not borne out in practice, and that the role of “pure protest” motives is very limited. Rather, the Extreme Right vote is driven by intense feelings of anti-immigrant sentiment in all countries but Italy. In line with theories of ethnic group conflict, these feelings are particularly strong within those segments of the electorate that compete with immigrants for scarce resources (low paid jobs and welfare benefits).

While the effects of anti-immigrant sentiment are strong, they are, however, moderated by general ideological preferences and party identification. On the basis of a new data set and a richer statistical model, these findings therefore confirm earlier claims that the Extreme Right vote can be explained by general causal mechanisms that apply to other parties, too (van der Brug and Fennema 2003). More specifically, the Extreme Right vote can be understood as the result of long-term political preferences and affiliations on the one hand and (immigration) policy-related attitudes on the other.21 Once these standard variables are measured adequately, it seems largely unnecessary to consider static22 and idiosyncratic factors like personality traits (Adorno et al. 1950) or alienation in today’s mass society (Kornhauser 1960). Rather, comparative electoral research should focus on the specific circumstances under which immigration is politicised and perceived as a problem that can move votes.

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1Endless debates not withstanding, there is still no agreement as to what is the most appropriate terminology. In practice, however, this has not hampered scientific progress. As Mudde (1996: 233) observes, “we know who they are, even though we do not know exactly what they are”. In the remainder of this paper, I shall use “Extreme Right” as a shorthand for the Austrian Freedom Party, the Flemish Vlaams Blok/Vlaams Belang, the French-speaking Belgian Front National, the Danish People’s Party and the Danish Progress Party, the French Front National and the Mouvement National Républicain (MNR), the Italian Alleanza Nazionale, Lega Nord and Movimento Sociale-Fiamma Tricolore, the Dutch Lijst Pim Fortuyn (LPF), and the Norwegian Progress Party, simply because it is the most common label for these parties.

2An attempt at a slightly stricter definition of the Extreme Right would involve three elements: i) while their economic policies are quite flexible and of lesser importance, parties of the Extreme Right take a tough stand on immigration and do often (though not always) take a “right” position with respect to many other issues that form the authoritarian-libertarian dimension of political conflict, ii) in terms of political style and patterns of co-operation with other parties within their respective political system, they are usually not well integrated and present themselves as outsiders or radical alternative to the established parties and elites, and iii) although they may be “extreme” in these respects, they are not necessarily “extremist”, i.e. beyond the liberal-democratic pale (see Arzheimer 2008 for a more elaborate discussion of these issues). While this definition still leaves considerable room for interpretation, in reality there is hardly any disagreement amongst scholars as to which parties belong to the Extreme Right family (Mudde 1996).

3While the Swiss SVP is often considered as a party of the Extreme Right, Switzerland was excluded because its institutional structure is vastly different from other West European countries and because until recently, the transformation of the SVP was confined to the so-called “Zurich wing” of the party.

4While the Extreme Right in Germany is slightly stronger than in Spain or the UK, Germany had to be excluded from this analysis because of the very low number of self-confessed supporters of the Extreme Right in the German part of the ESS.

5Following a well-established convention, latent variables are represented by ovals in Figure 1. Observable variables are represented by rectangles.

6From the information in the ESS, a simplified version of the Goldthorpe scheme (which is widely used in comparative research) was derived.

7The ESS team provides a scale of educational attainment that greatly facilitates international comparisons.

8The latter two variables – single person households and having/not having a partner – reflect notions of social isolation that are prominent in the older literature on right-wing extremism. Church attendance and union membership are primarily included as controls for the effects of traditional West European cleavages (Lipset and Rokkan 1967) but can also be interpreted as indicators for social integration.

9The variable was coded as trichotomous: identification with a party of the Extreme Right vs. identification with some other party vs. no identification at all (the reference category).

10Assertions about causality in non-experimental settings are always problematic. However, while variables in block I (socio-demographics) can clearly have a causal effect on the attitudes in block II (via socialisation and other processes of attitude formation), it is difficult to conceive of a process through which attitudes would affect socio-demographics. Similarly, the vote cannot possibly have a causal effect on socio-demographics. A causal effect of past behaviour on present attitudes via some sort of cognitive rationalisation process cannot be ruled out completely, though it seems unlikely that this would be a huge problem here.

11All models were estimated with MPlus 4.0, which provides estimators for both logit and probit links. Here, the latter was chosen because it is computationally much more attractive.

12Throughout this paper, the conventional five percent threshold is used.

13In Austria, the sign is correct but the effect is rather weak (though statistically significant). In Denmark, Italy, the Netherlands and in Norway, the parameter is not significantly different from zero. In Belgium and France, there is a weak but statistically significant effect that has the wrong sign.

14Because the number of Fiamma Tricolore and Lega Nord voters is very small (13), it is not possible to differentiate between them and the Alleanza voters.

15For simplicity’s sake, the other independent variables can be ignored since their effects are not significantly different from zero.

16The results refer to Norway but would be broadly similar for other countries. The values of 4, 7 and 5 for ideology reflect the lower quartile, upper quartile and median of the empirical distribution.

17The overall probability of a Freedom Party vote is rather high. This reflects the fact that the Freedom Party attracted more than 20 per cent of the vote in the Storting election of 2005.

18Empirically, the number of left-leaning, pro-immigrant Freedom Party identifiers is of course rather limited.

19The two latent variables are scaled so that a value of 0 is equivalent to the national average (see section 2). A value of +/-1 is one standard deviation above/below the national average.

20The calculations for Table 10 are based on the estimate for the respective coefficient in Table 9 (-0.035). However, the t-test test indicates that there is insufficient evidence (at the five per cent level) to reject the hypothesis that the coefficient is exactly zero. If one is willing to take the result of the test at face value, the ratio in Table 10 would be exactly 1.

21Presumably, candidate orientations are important, too, but these can not be measured with the data at hand.

22The (somewhat crude) indicators for alienation/social integration that are included in the model – household size, marital status, church attendance and union memberships – display few substantial effects in Tables 7, 8 and 9. The ESS questionnaire (like most other surveys) contains no indicators for personality traits, but the very notion of a disposition that is stable over decades is difficult to reconcile with the fluctuations of Extreme Right support in Western Europe. For a more comprehensive discussion and test of the traditional explanations of right-wing support in Western Europe see Arzheimer 2008.

Working Class Parties 2.0? Competition between Centre Left and Extreme Right Parties

 

1 Introduction

1.1 The Rise of the Extreme Right and the Transformation of Western European Policy Spaces

Over the last three decades, parties of the “radical”, “populist” or “extreme” right have become an almost ubiquitous feature of Western European party systems. During this “third wave” (Beyme1988) of radical right mobilisation, preexisting parties modified their ideological profiles (e. g. the Austrian Freedom Party, the Swiss People’s Party, the Scandinavian Progress Parties), and many more completely new parties emerged. While some of them were nothing more than a flash in the pan (e. g. New Democracy in Sweden, see Taggart 1996), others found more durable electoral support. As of today, almost all Western European political systems had to adjust (at least for a couple of years) to sustained Extreme Right mobilisation.

Initially, many observers interpreted these developments as a throwback to the Extreme Right’s inter-war onslaught on democracy (e.g. Prowe1994). But soon it became clear that the more successful amongst these parties departed in a crucial way from the political stances of the interwar extreme right movements and parties. Following the highly successful strategy of the French National Front (Rydgren2005), they abandoned biological racism, hyper-nationalism, and open hostility towards liberal democracy and instead made immigration (or more specifically the influx of non-West Europeans into Europe) their main issue. For that reason, some authors branded the emerging new party family simply as “anti-immigrant” (e.g. Fennema1997Fennema and Pollmann1998van der Brug, Fennema and Tillie2000Bjørklund and Andersen2002Gibson2002Boomgaarden and Vliegenthart2007Art2011), whereas others disputed the “single-issue thesis” (Mitra1988Mudde1999) or argued for a more nuanced classification of subtypes (e. g. Kitschelt1995Fennema1997Mudde2007).

This is certainly not the right space to re-open the (largely fruitless) “war of words” (Mudde1996) that dominated the scholarly debate in the 1990s. Today, most scholars working in the field agree on a set of stylised facts that can be summarised as follows:

  • While there are important differences amongst the “new” parties on the right in terms of their political traditions, policy positions, and general political style, these parties also display important similarities that set them apart from the Centre Right. Therefore, they should be grouped into a single (if very heterogeneous) party family.
  • While some of these parties harbour extremists and many of them are highly critical of single aspects of liberal democracy (most prominently minority protection), very few of them pursue a transition to authoritarian rule.
  • Therefore, “Radical” or “Extreme” (as opposed to extremist) Right are convenient shorthands for this party family.1
  • Immigration of non-western European people into Western Europe is not the only, but the single most important issue for all members of this party family. Mobilisation against immigrants and immigration is crucial for their electoral success.

Moreover, there is broad agreement that the rise of the Extreme Right presents politicians in Western Europe with a set of formidable challenges. First and foremost, their electoral success raised important questions of legitimacy. Did a vote for the Extreme Right indicate a more general lack of trust in the elites, or even a rejection of the democratic system? Was there reason to fear new “shadows over Europe” (Schain, Zolberg and Hossay2002), i. e. a return to the confrontational and often violent politics of the 1920s and 1930s? Should the existing parties engage in a dialog with their challengers or just ignore them?

Second, like the emergence of Green and Left-Libertarian parties, the rise of the New Right signalled a fundamental change in the patterns of party competition and co-operation in most Western European countries. For much of the postwar period, party competition in Western Europe was chiefly organised along a single left-right axis that largely reflected conflicts about economic redistribution (Fuchs and Klingemann1989van der Brug1999). However, both issues of the “New Politics” and matters of citizenship and immigration were not primarily perceived as economic problems and were therefore not easily aligned with the old left-right-conflict. Consequently, two or three dimensions are required to reconstruct the policy spaces of most Western European democracies (Kitschelt19941995Warwick2002Cole2005Bornschier2010), making party competition more complex and equilibria less likely.2

Third, and perhaps closest to the hearts of politicians, the zero-sum nature of electoral competition implies that the emergence of a new party family will bring about losses for existing parties in terms of votes, seats and eventually even ministerial portfolios. But which parties would suffer most?

1.2 Competition between Centre Left and Extreme Right Parties

From the party family’s moniker, one might be tempted to assume that the Centre Right had most to lose from the emergence of the Extreme Right, at least if voters primarily care about issues: In a classical Downsian (1957) perspective, demand for right-wing policies is fixed at least in the short- and medium term, and – depending on party positions and voters’ ideal points – the entry of a new competitor would significantly reduce the vote share of the Centre Right parties. If voters behave in line with a directional model (Merrill and Grofman1999), the outlook for the Centre Right is even starker, as voters who disagree with their radical policies may still vote for the Extreme Right for tactical reasons.

Aggregate trends of electoral support of electorate support in 16 Western European countries from the six decades since the end of World War II seem to corroborate these arguments: While support for the right as a whole3 has been largely stable, Christian democratic parties have on average lost about five percentage points of their electorate support while the Far Right could increase their share of the vote by almost seven points (Gallagher, Laver and Mair2011, 301).

Accordingly, much of the political and academic debate has focused on the negative implications that the rise of these parties has had for Conservative, Christian Democratic, Liberal, and Agrarian/Centre parties (e.g. Mair2001, 71).4 But Green/New left parties are perhaps the only ones not affected by the Extreme Right’s ascendancy, as these party families appeal to very different demographics and occupy diametrically opposed positions in Western European policy spaces.5

Taking a more analytical approach, Kitschelt (19941995) argued almost 20 years ago that a shift of the “main axis of partisan competition” was underway that would pit the New Left against the Extreme Right and present the Social Democratic/Centre Left parties with a conundrum: They would lose many of their more liberal voters to the parties of the New Left because they did not adequately represent the issues of the “New Politics” (Flanagan and Lee2003). At the same time, the Extreme Right would seize a sizable fraction of the working class vote, because the Centre Left had allegedly lost touch with their traditional voter base Bale (2003, 70-74).

But why would working class voters turn to the Extreme Right? Historically, support for the post-war Extreme Right had chiefly come from the “petty bourgeoisie” of artisans, small shop-keepers and farmers that made up the lower strata of the middle classes. This constituency was authoritarian and staunchly anti-communist/anti-socialist.

Working class voters, on the other hand, were often embedded in a network of trade unions and similar intermediate organisations, held strong preferences for redistribution, and were firmly attached to traditional left parties. Even if many voters (and some of the rank-and-file members) of these parties expressed a healthy degree of working-class authoritarianism (Lipset1959), elites and opinion leaders within the traditional working classes were firmly committed to principles of equality and international solidarity. Therefore, the idea of a large-scale swing from the Centre Left to the Extreme Right would have looked rather far-fetched three or four decades ago.

Through twin processes of de-alignment (Dalton, Flanagan and Beck1984) and social change (Crouch1999), however, swathes of the (non-traditional) working class have become available for other parties than the traditional left. Moreover, the Extreme Right has modified its programmatic appeal considerably over the six decades since the end of World War II, thereby becoming more palatable for members of the working class.


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Figure 1: Kitschelt’s 1995 view of Western European party systems


Perhaps the most radical interpretation of these programmatic changes was developed by Herbert Kitschelt in a highly influential monograph (Kitschelt1995). Kitschelt argued that under conditions of economic globalisation, workers outside the public sector would develop a taste for free market policies. At the same time, they would remain authoritarian with respect to their socio-cultural attitudes. According to Kitschelt, catering for these twin demands was the electoral “winning formula” that fuelled the unprecedented successes of the French National Front and the Austrian Freedom Party during the 1980s and early 1990s. A similar argument was developed by Betz in his seminal monograph (Betz1994). Figure 1, which slightly simplifies the presentation in Kitschelt (1995), shows the respective policy positions of Social Democratic, old style “Welfare Chauvinist” and more modern “Radical Right” parties.

In hindsight, however, the Extreme Right’s flirt with “neoliberalism” – presumably not a very serious affair in the first place – proved short-lived and inconsequential (de Lange2007). Within a few years after the publication of Kitschelt’s book, many Extreme Right parties had gone all the way from vocal champions of neoliberalism to globalisation critics, and the allegedly outdated “welfare chauvinist” strategy that campaigns for a strong but ethnically exclusionary welfare state had gained a lot of currency in Far Right circles. Consequentially, Betz (2003) has altoghether abandoned the idea that the Extreme Right does seriously pursue a “neo-liberal” agenda or has done so in the past, while Kitschelt has modified his original ideas considerably (McGann and Kitschelt2005).

Moreover, more recent research (Arzheimer2009b) demonstrates that there is no working class demand for “neo-liberal” policies. Where both members of the working class and the petty bourgeoisie support the Extreme Right, they tend to disagree on economic policies and cast their vote because the salience of economic issues is low (Ivarsflaten2005).

But even if the mid-1990s accounts by Betz and Kitschelt were wrong in their diagnoses, they clearly identified a very important symptom: Since the early 1980s, the Extreme Right has undergone a process of “proletarization and (uneven) radicalisation” (Ignazi2003, 216). At least for the relatively successful parties (e. g. the Austrian Freedom Party, the Norwegian Progress Party and the French National Front), there is some evidence for a trend from electorates that were heterogeneous or centred around a core of voters from the petty bourgeoisie towards more working class-dominated constituencies (Beirich and Woods 2000Betz 2002Bjørklund and Andersen 2002Mayer 19982002Riedlsperger 1998Rydgren 2003; see Oesch 2008 for a comparative cross-sectional analysis of Austria, Belgium, France, Norway, and Switzerland).

This new pattern of class-voting in Western Europe is not based on long-standing party loyalties but rather on group- and policy-related attitudes: Public opinion data consistently shows that the Extreme Right vote is driven by intense worries about immigrants and immigration6 that are most prevalent amongst voters with low levels of educational attainment who are either unemployed or holding blue-collar jobs.7

While many authors frame these worries as “resentment” and interpret the underlying policy dimension primarily in terms of “culture” and “identity”, one should not ignore the fact that concerns about immigrants and immigration have clear economic underpinnings: The vast majority of immigrants in Western Europe are unskilled or semi-skilled workers. Obviously, members of the working class are much more likely to perceive these persons as an economic threat than middle class voters, who might actually benefit from the additional supply of cheap labour.


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Figure 2: An updated perspective on Western European party systems


On the whole, research since the mid-1990s suggests that patterns of party competition and class voting have indeed changed, although in a way that is quite different from Kitschelt’s original reading of the situation (see figure 2). Instead of converging on the “Radical Right” strategy, parties of the Extreme Right are looking for a (not very) “new winning formula” (de Lange2007) and have incorporated elements of “welfare chauvinism” into their manifestos, although to a varying degree. Social Democratic parties, on the other hand, have cautiously moved to more economically centrist (and arguably more socially liberal) positions in a bid to respond to the new challenges of the 21st century and to become more attractive for middle-class voters (see Keman 2011 for a comprehensive analysis that outlines the extent of this shift in 19 polities). This programmatic change opened up additional space for the Extreme Right and made it even easier for them to poach working class voters from the Centre Left. That raises the question whether there is anything the Centre Left can do about this development.

The remainder of this chapter is organised as follows. Section 2 gives a brief overview of the data base and the statistical models and methods used for its analysis. Section 3 presents a comparative longitudinal analysis of the “proletarisation” of the Western European Extreme Right Vote since 1980. Section 4 directly looks at the competition between Extreme Right and Centre Left parties for the working class vote. Finally, section 5 briefly summarises the findings.

2 Data, Model, Methods

 


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Figure 3: The spacing of relevant Eurobarometer surveys in time and across countries

 


The analyses presented in the following sections cover the member states of the European Union (EU) as it existed before the Eastern enlargement rounds, plus Norway. Survey data come from the Mannheim Eurobarometer Trend File (Schmitt et al.2009a,b), a partial cumulation of the bi-annual series of Eurobarometer surveys that greatly facilitates cross-national and longitudinal analyses. The temporal coverage of these data spans the whole period of the Extreme Right’s electoral ascendancy during the 1980s and 1990s, as well as a few years of the new millenium.

There are, however, a few gaps: Data for Austria, Finland, Sweden and Norway are not available for the whole period. Moreover, surveys without any supporters of the Extreme Right had to be excluded, which removed the United Kingdom and the Republic of Ireland from the analysis.8 Figure 3 gives a graphical overview of the spatial and temporal coverage.

Information on social class in the Eurobarometer series is effectively restricted to present occupation. To simplify the presentation, respondents were coded as holding blue-collar jobs (“workers”), belonging to the petty bourgeoisie (“farmers and owners”), holding any other occupation (“other”), being unemployed, or being retired.9

In order to model contextual effects on right-wing voting, the Eurobarometer surveys were augmented with macro data. Information on unemployment rates and unemployment benefits comes from the OECD (200220032004), while data on new asylum applications – in the Western European context, a very useful proxy for actual immigration figures – were taken from reports compiled by the OECD and the Office of the United Nations High Commissioner for Refugees (OECD1992UNHCR2002).

Finally, the Comparative Manifesto Project database was used to construct a series of five variables that capture the positions of mainstream parties with respect to the issues of the Extreme Right, i.e. “internationalism”, “multi-culturalism”, “national lifestyle”, and “law and order” (see Arzheimer and Carter (2006); Arzheimer (2009a) for a more detailed discussion of the rationale behind these measures). These variables pertain to the position of the respective Social Democratic party, the most extreme position taken by any other mainstream party, the salience of these issues for the Social Democrats, the salience for all other mainstream parties, and the variation in policy positions across all other mainstream parties.10

To account for the hierarchical nature of the data (respondents are nested within 336 survey waves that were conducted in 15 polities), binary logistic multi-level models are specified. Because the Extreme Right is persistently stronger in some countries (e. g. Belgium and France) than in others (say Spain and Germany), stable unit (country) effects are represented by a series of dummies.11 These dummies are also required to control for changes in the national composition of the sample over time. Specifying country effects leaves just two levels of analysis: voters, and the particular contexts in which they were interviewed.

Even when controlling for unit effects and contextual information, it makes sense to assume that people who are interviewed in the same survey wave are subject to common random political shocks that affect their voting behaviour. These shocks are modelled as draws from a Gaussian distribution with standard deviation σu, which estimated from the data in addition to the usual parameters. As a result of these shocks, respondents in the same context will give more similar answers than one expect by chance alone. The intraclass correlation coefficient ρ which ranges from 0 to 1 is a measure for this similarity, with values closer to unity indicating greater alikeness within a context.12

All models were estimated using the xtlogit procedure in Stata 11.2. Checks indicate that the number of quadrature points used was sufficient to guarantee stable estimates.

3 The Proletarisation of the Western European Extreme Right Vote, 1980-2002

The idea of a “proletarisation” (Ignazi2003) of the Western European Extreme Right features prominently in the literature, but very little comparative cross-temporal empirical evidence for this alleged development has been presented so far. With the Eurobarometer Trend File, however, it is possible to trace the purported trajectory of the Extreme Right’s electorate.

 


Fixed country effects omitted

(1) (2)
Worker 0.483∗∗∗ 0.441∗∗∗
(0.0277) (0.0307)
Farmer/Owner 0.438∗∗∗ 0.478∗∗∗
(0.0347) (0.0363)
Retired 0.0546 0.0563
(0.0282) (0.0318)
Unemployed 0.555∗∗∗ 0.552∗∗∗
(0.0410) (0.0455)
Time 0.00593∗∗∗
(0.000666)
Worker × Time 0.00176∗∗∗
(0.000433)
Farmer/Owner × Time -0.00207∗∗∗
(0.000512)
Retired × Time -0.0000549
(0.000442)
Unemployed × Time 0.000120
(0.000665)
Observations 254726 254726
σu 0.720 0.621
ρ 0.136 0.105
Groups 336 336
t statistics in parentheses
∗ p< 0.05  , ∗∗ p <0.01  , ∗∗∗ p <0.001

Table 1: Sociodemographic factors and the extreme right vote, 1980-2002/3


The left column (1) of table 1 shows the estimates from a simple socio-demographic multi-level model of Extreme Right voting in Western Europe. The model is based on just under 255000 interviews.

As can be seen from the coefficients, being unemployed or belonging to the working class or the petty bourgeoisie considerably increases the chances of an extreme right vote, compared to the “other” category. Either factor increases the logit of an Extreme Right vote by 0.4 to 0.5 points. Being retired, on the other hand, does not make an appreciable difference.

The exact impact of this increase depends on the fixed country effects but is roughly proportional to a 50 per cent change in the probability of the Extreme Right vote. In Austria, for instance, members of the “other” group have an estimated probability of just under 15 per cent of voting for the Freedom Party. For workers, the estimated probability is almost 22 per cent.

The term proletarisation, however, implies change over time. In the right column (2) of table 1, the membership indicator were interacted with an additional variable that represents the time (in months) at which the survey was taken. In order to minimise collinearity, the variable was centred so that it takes a value of zero for March 1991, which is the midpoint of the period under observation. Given the huge range of the time variable (see table 2), it is not surprising that the estimated coefficients are very small. Nonetheless, the picture that emerges is remarkably clear. The effect of being a pensioner is essentially stable, while the effect of being unemployed increases only very slightly over time. The effect of being a member of the working class, on the other hand, becomes considerably stronger with time, while the effect of belonging to the petty bourgeoisie becomes weaker at roughly the same rate.

Taken together, these results show that the Extreme Right electorates indeed underwent a process of proletarisation between 1980 and the early naughties. Moreover, these findings cannot be ascribed to changes in the composition of the sample (i.e. the accession of Greece, Spain and Portugal to the European Union during the 1980s and the 1995 enlargement), because fixed country effects are controlled for. Therefore, the interaction effects represent common trends across all 15 polities. This constitutes the first truly comparative and longitudinal evidence for a general proletarisation of the Extreme Right vote in Western Europe.

But how important are these trends in substantive terms (i.e. votes and seats)? Again, the exact size is context-dependent and most easily illustrated by calculating estimates for an arbitrary country. The estimated vote share of the Danish Extreme Right amongst workers in 1980, for instance, was just under two per cent, while the respective figure for members of the Danish petty bourgeoisie was about three per cent. In 2002, the estimate for the petty bourgeoisie was eight per cent, while the figure for the working class has risen to almost 13 per cent. Although the Extreme Right has made considerable inroads into both groups, the ratio of the respective propensities to vote for the Extreme Right has been reversed. Therefore, it makes indeed sense to talk about a proletarisation of the Extreme Right vote. This trend is further amplified by the fact that the petty bourgeoisie is shrinking even faster than the working class.

One should, however, not throw out the baby with the bath water: Precisely because the working class is in decline, there is a natural limit to this process. Moreover, while social class has obviously lost some of its previous importance (Clark, Lipset and Rempel1993Nieuwbeerta and Graaf2001), its effect on the probability of voting for the traditional left has by no means disappeared completely (Evans2001). Thus, the next section section will look specifically at the competition between Extreme Right and Social Democratic parties over the working class vote.

 


min p25 mean p75 max
XR vote 0.00 0.00 0.04 0.00 1.00
Worker 0.00 0.00 0.18 0.00 1.00
Farmer/Owner 0.00 0.00 0.10 0.00 1.00
Retired 0.00 0.00 0.22 0.00 1.00
Unemployed 0.00 0.00 0.06 0.00 1.00
Time -131.00 -36.00 10.22 56.00 130.00
AT 0.00 0.00 0.03 0.00 1.00
BE 0.00 0.00 0.07 0.00 1.00
DE-E 0.00 0.00 0.06 0.00 1.00
DE-W 0.00 0.00 0.14 0.00 1.00
DK 0.00 0.00 0.14 0.00 1.00
ES 0.00 0.00 0.03 0.00 1.00
FI 0.00 0.00 0.03 0.00 1.00
FR 0.00 0.00 0.11 0.00 1.00
GR 0.00 0.00 0.06 0.00 1.00
IT 0.00 0.00 0.12 0.00 1.00
LU 0.00 0.00 0.01 0.00 1.00
NL 0.00 0.00 0.13 0.00 1.00
NO 0.00 0.00 0.03 0.00 1.00
PT 0.00 0.00 0.02 0.00 1.00
SE 0.00 0.00 0.02 0.00 1.00
N  254726

Table 2: Sociodemographic model: summary statistics


4 Left or (Extreme) Right? The Western European Working Class Vote, 1980-2002

In their recent analysis of Social Democratic reactions to the rise of the Extreme Right, Bale et al. (2010) have usefully identified three elements of this challenge, and three strategies available to the Centre Left: The presence of Extreme Right parties will heighten the salience of “right” issues in general, can increase the number of potential coalition partners for the Centre Right, and may lure working class voters away from the left. Social democratic parties can respond by holding on to their traditional relatively tolerant position towards immigrants, by trying to “defuse” the immigration issue, or by shifting their position (Bale et al.2010, 412).

As Bale et al. (2010, 413-414) point out, the effectiveness of the “defuse” strategy is very limited, making the first strategy the default, as Social Democratic party elites are normally committed to values of tolerance and international solidarity. Therefore, they will find it difficult to abandon their support for relatively liberal immigration policies to avoid political losses. Such normative convictions seriously restrain the Centre Left’s room for manoeuvre.


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Figure 4: Ideological movement of Social Democratic parties over time


Nonetheless, the qualitative analysis of developments in Austria, Denmark, the Netherlands and Norway by Bale et al. shows that Social Democratic parties have sometimes modified their positions on the immigration dimension (see Bale et al.2010, 421 for an overview). A quantitative analysis (see figure 4) of the CMP-Data provides further evidence for such programmatic shifts: Although there is considerable national variation, Social Democratic parties in many countries including Germany, Denmark, Finland, France, Italy, and the Netherlands have taken consistently tougher stands on issues of migration and national identity over the years.

 


Fixed country effects omitted

(1) (2) (3)
Male 0.445∗∗∗ 0.449∗∗∗ 0.448∗∗∗
(0.0515) (0.0517) (0.0517)
Time 0.00982∗∗∗ 0.00692∗∗∗ 0.00651∗∗∗
(0.000874) (0.00121) (0.00127)
Toughness (max SD) 0.0327
(0.0270)
Toughness (mean SD) 0.0296
(0.0309)
Ideology Salience (SD) -0.0437 -0.0383
(0.0257) (0.0247)
Toughness (other) -0.00246 0.00360
(0.0255) (0.0242)
Ideological Variance (other) -0.0131∗∗ -0.0137∗∗
(0.00437) (0.00429)
Ideology Salience (other) 0.119∗∗∗ 0.116∗∗∗
(0.0291) (0.0288)
New Asylum Applications 0.0386 0.0326
(0.0667) (0.0663)
Unemployment 0.0999∗∗ 0.106∗∗
(0.0374) (0.0388)
Replacement Rate 0.0515∗∗∗ 0.0520∗∗∗
(0.0138) (0.0138)
Observations 19858 19663 19663
σu 0.733 0.645 0.646
ρ 0.140 0.112 0.113
Groups 336 327 327
t statistics in parentheses
∗ p< 0.05  , ∗∗ p <0.01  , ∗∗∗ p <0.001

Table 3: Full model: XR vs. Social Democratic vote amongst working class respondents


But how do working class voters respond to this repositioning of the Centre Left? The left column (1) in table 4 gives the estimates for the coefficients of a very simple baseline model. The sample is restricted to working-class respondents who intend to vote either for a Social Democratic party (0) or and Extreme Right party (1). The model features a single sociodemographic control to account for the well-known gender gap, and a linear (in the logits) trend factor. Like the models in the previous section, the model also contains fixed country effects to account for stable differences between polities. Estimates for these effects (not tabulated) are very low in countries as diverse as Germany (-3.3), Spain (-6.3), Finland (-4), Luxembourg (-4.6), Portugal (-5.7), or Sweden (-5.2), which implies that in these countries, the odds of a Social Democratic vote are between 27 (exp(3.3)) and 545 (exp(6.3)) times higher than the odds of an Extreme Right vote.

There is, however, a set of countries including Austria (-1.7), Belgium (-2), Denmark (-2.2), France (-2.4), and particularly Italy (-.65), where the odds of an Extreme Right vote are much higher in comparison. While the result for Italy might be due to the fact that the AN as the largest relevant party in the country has become relatively moderate since the 1990s, the findings for the other countries are striking: Across the board, a Social Democratic vote is only between 5.5 and 11 times more likely than an Extreme Right vote in this core constituency of the Centre Left.

Moreover, the trend factor indicates that the odds of an Extreme Right vote have risen considerably over time: If one is prepared to take the model estimates at face value, the odds of a working class respondent voting for the Extreme Right increases by a factor of almost 13 (exp(0.0098 × 261)) between the first and the last survey wave. Even if one takes potential deficiencies of the data and model specification into account, this clearly demonstrates that Social Democratic parties are losing support amongst working class voters.

While this is certainly an interesting finding in itself, time is chiefly used as a control in a second series of models (columns (2) and (3)) that build on Arzheimer’s (2009a) contextual model of Extreme Right voting.13 This amended model allows for a direct test of the viability of two of the strategies outlined by Bale et al. as well as for an indirect test of the third.

Since some elections were contested by two or more parties that were classified as Social Democratic by the CMP, Social Democratic ideology was operationalised in two variants: “Toughness” refers either to the most right-leaning party (column (2)) or to the average of all Social Democratic party positions, weighted by the respective party’s share of the vote (column (3)).14 However, the way Social Democratic ideology is measured makes virtually no difference.

According to this second set of estimates, the trend towards more Extreme Right voting is slightly less pronounced15 once the additional contextual variables are taken into consideration. Nonetheless, given its wide range time still has the strongest effect amongst all covariates.

The level of welfare state protection as measured by the OECD’s standardised wage replacement rate for the unemployed also has a strong positive effect on the probability of an Extreme Right vote. Raising the standards from the first to the third quartile of its empirical distribution (see table 4) will almost quadruple the odds of a right-wing vote. Given the Extreme Right’s rediscovery of centre-left leaning policies, this could be interpreted as a result of “welfare chauvinism” and (perceived) ethnic competition (Bélanger and Pinard1991) over a resource that is still plentiful. However, an alternative explanation is at least as plausible: Only if the welfare state is seen as safe and can be taken for granted, workers will turn from Social Democratic parties towards the Extreme Right.

Another factor that has a strong effect on the electoral prospects of the Extreme Right is the salience of their issues for other parties (excluding the Social Democrats). The more statements other parties make on questions of immigration, national identity and the like, the better the Extreme Right does in the polls, irrespective of the direction of these statements. Since objective factors such as unemployment and new asylum applications (which have weak or insignificant effects) are statistically controlled for, this finding can be interpreted as evidence for an agenda setting effect (Arzheimer2009a).

Ideological variation in the manifestos of other parties has a moderate negative effect on right-wing voting, whereas ideological “toughness” (i.e. attempts by mainstream parties to steal the immigration issue) does not shift the balance between the Extreme Right and the Social Democrats.

Taken together, the effects of salience and ideological variation indicate that a strategy of issue diffusion could be viable in principle, if (and only if, as the Social Democrats can hardly shape political discourse singlehandedly) the other mainstream parties co-operate.

 


min p25 mean p75 max
XR vote 0.00 0.00 0.12 0.00 1.00
Male 0.00 0.00 0.60 1.00 1.00
Time -131.00 -47.00 1.99 55.00 130.00
Toughness (max SD) -11.71 -2.01 -0.12 1.51 13.68
Toughness (mean SD) -11.71 -2.37 -1.02 1.12 7.45
Ideology Salience (SD) 0.00 3.45 6.83 9.19 16.08
Toughness (other) -4.54 0.59 4.84 7.92 27.54
Ideological Variance (other) 0.00 1.87 17.18 16.50 244.60
Ideology Salience (other) 0.50 5.08 8.95 12.41 31.25
New Asylum Applications -0.98 -0.61 0.16 0.58 4.46
Unemployment -4.91 -1.31 0.35 1.69 12.29
Replacement Rate -31.62 -4.19 4.07 18.48 32.96
AT 0.00 0.00 0.05 0.00 1.00
BE 0.00 0.00 0.06 0.00 1.00
DE-E 0.00 0.00 0.06 0.00 1.00
DE-W 0.00 0.00 0.19 0.00 1.00
DK 0.00 0.00 0.17 0.00 1.00
ES 0.00 0.00 0.03 0.00 1.00
FI 0.00 0.00 0.02 0.00 1.00
FR 0.00 0.00 0.12 0.00 1.00
GR 0.00 0.00 0.04 0.00 1.00
IT 0.00 0.00 0.05 0.00 1.00
LU 0.00 0.00 0.00 0.00 0.00
NL 0.00 0.00 0.10 0.00 1.00
NO 0.00 0.00 0.05 0.00 1.00
PT 0.00 0.00 0.04 0.00 1.00
SE 0.00 0.00 0.01 0.00 1.00
N  19663

Table 4: Full model: summary statistics


While this test of the “defuse” strategy might be somewhat indirect, the efficiency of the “hold” and “adopt” strategies can be more readily assessed by looking at the estimates for the “toughness” and salience variables that refer to Social Democratic parties. Neither of them has a significant effect on the odds of voting for the Extreme Right. Put differently, in this core constituency of the Centre Left, it does not make a difference whether the Social Democrats stick to their traditional positions on immigration or whether they try to toughen up their policies. Either way, their fortunes vis-a-vis the Extreme Right are largely determined by external factors and an overall negative trend.

The null effect of salience provides an interesting correlate. This variable takes a value of zero if Social Democrats completely ignore the issues of the Extreme Right, which is equivalent to a very radical “defuse” strategy, whereas positive values represent attempts to engage with the issue by making affirmative and/or critical statements. The insignificance of the coefficient provides further evidence for the assertion that a “defuse” strategy is only viable if pursued in concert.

5 Conclusion

After World War II, parties and movements of the Extreme Right were most closely associated with the petty bourgeoisie. Over the last three decades, however, the propensity of workers to vote for the Extreme Right has risen significantly. This “proletarisation” is the result of the interplay between a long-term dealignment process and increasing worries amongst the European working classes about the immigration of cheap labour. As a result, Western European Centre Left parties may find themselves squeezed between the New Right on the one hand and the New Left on the other.

The analyses in the previous section have shown that there is no obvious strategy for dealing with this dilemma. Staying put will not win working class defectors back. Toughening up immigration policies is unpalatable for many party members, does not seem to make Social Democrats more attractive for working class voters, and might eventually alienate other social groups.

That leaves what Bale et al. have called the “defuse” option, i.e. efforts to downgrade the immigration issue. In democracies, however, a single party can not normally sustain control over the political agenda. Any attempt to de-politicise immigration would therefore require some sort of agreement amongst mainstream parties. Given that Centre Right (Bale2003) and (for completely opposite reasons) even New Left parties might have a strategic interest to keep the debate on immigration alive, this is not a very likely outcome. In all probability, the working class parties “of a new type” will keep poaching voters from the Social Democrats.

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1I will treat these two terms as interchangeable through the remainder of this chapter.

2For a slightly different account of these developments see van der Brug and van Spanje (2009), who claim that European parties’ actual policy proposal can still be arranged on a single vector even though parties and voters operate in a two-dimensional space.

3Gallagher, Laver and Mair subsume five party families under this label: Christian democrats, Conservatives, Liberals, Agrarian and Centre parties, and the Far Right.

4Other authors, however, have highlighted the strategic opportunities that the rise of the new party family may present for the right as a whole if and when the Extreme Right can be brought into a coalition (Bale2003).

5Consequently, the rise of the Extreme Right has sometimes been framed as a “silent counter-revolution” (Ignazi1992) against the growing influence of the New Left and their post-materialist electoral base.

6These feelings are related to, but not identical with xenophobia and racism (Rydgren2008).

7See e. g. van der Brug, Fennema and Tillie (2000) and Arzheimer (2009b) for reviews of the importance of ideology and Arzheimer and Carter (2009) for the nexus between class and attitudes.

8The OECD does not provide Standardised Unemployment Rates for Luxembourg. Thus, the country had to be excluded from the series of models presented in section 4.

9Homemakers were coded according to the occupation of the householder, if available.

10For the construction of the two latter variables, positions were weighted with the parties’ shares of the vote. In some cases, elections were contested by two or more parties codes as Social Democratic by the CMP. See section 4 for details.

11East and West Germany are treated as two separate polities.

12ρ equals the proportion of total variance contributed by σu.

13To ease the estimation and interpretation, a number of interaction effects and relatively stable macro variables were dropped. Moreover, all attitudinal and most socio-demographic variables were dropped, since they do not vary much in this subset of working class voters. The findings for many variables are somewhat different from those reported in Arzheimer (2009a) because they apply to a more limited choice set and a subsample of the original data.

14The salience variable was always constructed as an weighted average over all Social Democratic party positions in the respective election (if applicable).

15The estimated factor change in the odds is exp(0.007 × 261) = 6.

How (not) to Operationalise Subnational Political Opportunity Structures

 

Analysing the influence that features of the national political context exert on the vote for Western Europe’s ‘extreme’ or ‘radical’ right[1] parties is now a minor industry (see for example Arzheimer & Carter 2006; Golder 2003; Jackman & Volpert 1996; Knigge 1998; Lubbers et al. 2002; Swank & Betz 2003), but in a recent contribution to this journal, Kestilä and Söderlund (2007) argue that research should give more consideration to what they call ‘subnational political opportunity structures’. According [k1] to Kestilä and Söderlund, focusing on the subnational context within one country mitigates against three problems that trouble the existing contextual analyses: i) at the subnational level, the number of contexts is large in comparison to the number of relevant variables; ii) unique features of the party system are obviously held constant; and iii) the heterogeneity of the radical right party family need not be of concern (2007: 774-775). [k2] This theoretical claim is backed up by an ecological analysis of the French regional elections of 2004. In a straightforward linear regression at the level of the département, Kestilä and Söderlund relate the electoral support for the Front National (henceforth FN) in the first round of those elections, as well as an index of the FN’s electoral success (that assesses the FN’s success in relation to the leading contender), to five aggregate variables: turnout in the first round of the 2004 election, the logged district magnitude in the previous regional elections of 1998, the effective number of party lists (Laakso-Taagepera index) in 1998, the share of immigrants born outside the European Union in 1999, and the unemployment rate in 1999. They find that turnout and district magnitude have significant negative effects on the FN’s electoral support,[2] whereas the effects of the number of party lists and unemployment are positive and significant. Most interestingly, the effect of immigration is not significantly different from zero in Kestilä and Söderlund’s model of FN aggregate support. From these results, Kestilä and Söderlund conclude that the radical right benefits from low turnout levels, and that greater proportionality of the electoral system does not increase support for the radical right but is actually related to substantially lowerlevels of support. They also conclude that the FN benefited when the effective number of party lists in 1998 was high, and when unemployment levels were high. By contrast, the share of immigrants present in each département did not affect the FN’s electoral score.

 

It should be noted that, while the results that relate to the district magnitude are in line with the findings of some other studies (e.g. Arzheimer & Carter 2006), those that pertain to unemployment and immigration are not: in country-level analyses the effect of unemployment is subject to on-going discussion, and the effect of immigration has been consistently found to be strong and positive.

 

From their results, and the findings of other studies notwithstanding, Kestilä and Söderlund argue that the ‘subnational political opportunity structure has been of great importance for the FN’ and more generally, that the subnational approach ‘is able to control a wider range of factors pertaining to the political system and tends to provide more reliable results’ (2007: 790).

 

Kestilä and Söderlund have invited other scholars working in the field to engage in a discussion of their unexpected findings. We have taken up this invitation because, while we concur that features of the subnational context are potentially relevant for the radical right vote and should be incorporated into more comprehensive accounts of support for these parties, we are not convinced by Kestilä and Söderlund’s conceptualization of what constitutes a subnational political opportunity structure, and nor are we persuaded by the empirical evidence they present.

 

We begin this article by highlighting some of the theoretical, conceptual and methodological problems present in Kestilä and Söderlund’s study. Then, since Kestilä and Söderlund were extremely forthcoming in providing us with their data, we engage in some re-analysis. In doing this we discuss the difficulties of estimating and interpreting the coefficients in Kestilä and Söderlund’s model and, using an indicator for the FN’s regional entrenchment that is independent of district magnitude and of the effective number of party lists, we demonstrate that features of the subnational political opportunity structure included in Kestilä and Söderlund’s model are essentially spurious. We close by offering an alternative operationalization of one of the key variables contained in Kestilä and Söderlund’s model.

 

 

Subnational Political Opportunity Structures: Conceptual Difficulties and Operationalization

 

The concept of political opportunity structures is notoriously vague, but at its core is the idea that certain variables can capture the degree of ‘openness or accessibility of a political system for would-be political entrepreneurs’ (Arzheimer & Carter 2006: 422). If one accepts this as a working definition, it follows that subnational political opportunity structures refer to a set of regional or local conditions that would either facilitate or hamper the attempts of the radical right to mobilize voters. Precisely because the concept of political opportunity structures is so vague, identifying these conditions and operationalizing these variables is a tricky task, and unfortunately, there are problems with the way in which Kestilä and Söderlund have gone about this in their study.

 

Our first misgiving concerns the inclusion of (regional) party system fragmentation in Kestilä and Söderlund’s model, operationalized by the variable ‘effective number of party lists’. As Kestilä and Söderlund suggest, the level of party system fragmentation might indeed be important to small or new parties either because a high level of fragmentation at the previous election might indicate that the system is open and therefore more favourable to such parties or, conversely, because a high level of fragmentation might indicate that a wide variety of alternatives already exists, rendering it difficult for a small or new party to make a breakthrough (2007: 784).

 

However, using this variable in relation to the FN is problematic because, as Kestilä and Söderlund themselves point out, the FN is neither small nor new: it is a well established political competitor that has acquired the status of third political force in many parts of France (2007: 775). Therefore, given that it is not knocking on the door of the French party system but is already clearly inside it, the issue for the FN is not how accessible the party system is (which is what party system fragmentation measures), but is how much political space the party has or, put differently, how much competition it faces at the right end of the political spectrum. In short, we would argue that party system fragmentation is not an appropriate variable with which to measure party competition in this instance and that a much more relevant variable would be one that taps the ideological space available to the FN (see below).

 

Kestilä and Söderlund do briefly discuss the ideological aspect of party competition when they note that cross-national studies that have analysed the impact of political opportunity structures have had to operationalize and measure ideological convergence or divergence and issue adoption. However, although they recognize ‘that the local party organisations may have an agenda somewhat deviant from the national one’ (2007: 783), they do not include the ideological dimension of party competition in their model on the basis that a subnational analysis such as theirs benefits from being able to ‘hold the ideological differences constant in each subunit due to the national character of the campaign in the elections of 2004’ (2007: 775).

 

Now, it might indeed be the case that the campaign for the regional elections of 2004 took on a national character, but that does not mean that the contest was ideologically similar in all départements. A quick glance at the first round of the 2004 regional election results reveals that in some regions competition on the right of the political spectrum was played out simply between the FN and one mainstream right party list (for example in Picardie). Elsewhere, however, there was more than one mainstream right party list and/or more than one list from the radical right. In Aquitaine, for instance, there were two mainstream right party lists, while in the Rhône-Alpes region not only was there an FN list, but there was also another ‘extreme right’ list. Similar patterns can be found with respect to the 1998 regional elections – the results of which Kestilä and Söderlund use to calculate their ‘effective number of party lists’.

 

The ideological nature of party competition has therefore varied by region and voters have been faced with a different choice of party lists according to where they live. This is likely to be important in an explanation of the FN’s success as the party may well be hindered by the presence of multiple mainstream right lists, and may also experience lower vote shares in départements where an alternative radical right list exists. By only including the party system fragmentation variable in their model, Kestilä and Söderlund fail to account for these trends.

 

Including the effective number of parties or party lists in analyses of the radical right party vote is also problematic for methodological reasons. This is because the vote share of the very party whose electoral success is being explained (in this case the FN) is included in the calculation of the effective number of parties: (1 / S pi2) for N parties, where p is the vote share of party i). This means that the two variables cannot possibly be independent of each other, and that, given the construction of the index, there must be a non-linear relationship between them, the exact shape of which depends on both the number and relative strength of other parties.

 

The co-variance of these two variables is most easily illustrated by simulation. To do this, we selected the results from four random départements in the 1998 regional elections (since départements were treated as districts in 1998) and let the FN’s vote share vary around its empirical value while holding the relative support within the mainstream right bloc and the absolute support for all other party lists constant.[3] Figure 1 illustrates the results of this simulation and clearly shows that, at least in these four départements, a change in the fortunes of the FN within the département would ceteris paribus have a strong, positive and almost linear impact on the effective number of parties.[4]

 

 

[Figure 1 about here]

 

 

By including the effective number of party lists in 1998 in their model, Kestilä and Söderlund therefore effectively regress the FN’s success in 2004 on a variable that already encompasses previous levels of support for the party. This is problematic for theoretical reasons, and our simulation shows that it also has very real implications in terms of interpreting the effect of this variable.

 

Turning to district magnitude, we take no issue with Kestilä and Söderlund’s decision to include this variable in their model. For theoretical reasons it makes sense to do so: it will be harder for the FN – a medium-sized party – to win votes in districts with a small magnitude than it will be for it to be successful in districts with a greater magnitude. Moreover, this effect will be exacerbated if the number of potential voters is small to begin with and if the party decides to invest fewer resources in districts with a small magnitude than it does in those with a larger magnitude.

 

What we are unhappy with, however, is the way in which this variable has been operationalized in Kestilä and Söderlund’s study. They regress the FN’s vote in each département in 2004 on the (logged) district magnitude in the 1998 regional election, and we would argue this is troublesome for two reasons.

 

Firstly, the inclusion of the (logged) district magnitude ignores the effect of legal thresholds which can and do override the effects of district magnitude. In the case of the electoral system of 1998, the five per cent legal threshold in place at the département level effectively cancelled out the effects of district magnitude in départements with a magnitude of 14 or more. Given that the district magnitude was 14 or more in 50 of the 94 départements, this has large implications for the model, and, as such, it would have made greater sense to include the effective magnitude or the effective threshold rather than simply the (logged) district magnitude.[5]

 

The second reservation we have about Kestilä and Söderlund’s operationalization of district magnitude concerns their decision to use the (logged) district magnitude in the 1998 regional election – i.e. in the previous regional election. Kestilä and Söderlund do this because they maintain that ‘changes in electoral laws may not necessarily have an immediate effect’ and that the psychological effect of an electoral system may take a while to manifest itself. They also note that ‘the district magnitude of 1998 and seats allocated to departments in 2004 have a very strong correlation’ (2007: 792, note 7).

 

We certainly do not dispute the fact that psychological effects of electoral systems may take a while to register with voters, and as such, other than keeping an eye on our comments above about legal thresholds, we would have no criticisms of the use of this variable had the electoral system of 1998 been identical to that used in 2004. The problem, however, is that it was not: the electoral system used in 1998 was fundamentally different to that used in 2004.

 

The system used in the 1998 regional elections (in use since the first regional elections of 1986) was a one-round proportional electoral system. Between 3 and 72 seats were distributed at the level of the département and, as mentioned above, there was a five per cent legal threshold in place. In 2004 a new two-round electoral system came into operation, however. Under this system, even though seats were eventually divided up between departmental sections, it was at the level of the region that lists were presented, votes were aggregated and seats were distributed. The district magnitude of the regions ranged from 43 in the Limousin and Franche-Comté regions to 209 in the Ile-de-France region but the existence of legal thresholds meant that the effect of district magnitude was effectively cancelled everywhere.[6]

 

Given that the electoral system changed so fundamentally between 1998 and 2004, we believe that it is unrealistic to argue that the 1998 system still exerted a psychological effect on voters and political elites in 2004. If voters are well-informed and rational enough to react to the mediating effects of electoral systems in the first place, then they are hardly likely, on the one hand, to take the effects of the 1998 electoral system into account, and yet, on the other, to fail to notice that the system has been changed thoroughly in the interim. And as concerns political elites, the effects of the 1998 system will not have entered into their calculations in 2004. Rather they will have taken the new electoral system into account when they decided on their campaigning strategies and on the resources they would invest in each district for the 2004 contest.

 

For these two reasons, therefore, we would argue that it does not make sense to use the (logged) district magnitude in the 1998 regional election as an indicator of the openness or accessibility of the political system in 2004. And the fact that the district magnitude of 1998 correlates very strongly with the seats allocated to départements in 2004 does not allay our fears because this correlation does not take account of the effects of legal thresholds and because the number of seats distributed to départements in 2004 is irrelevant since the allocation of seats took place at the level of the region, not at that of the département.

 

We are also uneasy about the inclusion of turnout in Kestilä and Söderlund’s model. The issue here is not how this variable has been operationalized, but rather why it is included in the model at all.

 

Of course, turnout is commonly included in national and comparative election studies, and a number of these works have observed a negative correlation between turnout and support for parties that are not fully integrated into the party system (Reif et al. 1997; van der Eijk et al. 1996), including the FN, whose vote share has been found to be highly correlated with turnout in both presidential and legislative elections (Auberger 2008). It makes good sense to include turnout in studies of this kind since they are in the business of explaining patterns in individual voter behaviour. In this instance, they are able account for the negative correlation they find by arguing that, while politically dissatisfied supporters of the established parties may refrain from voting altogether, politically dissatisfied supporters of non-established parties can express their dissatisfaction with their vote.

 

The purpose of an analysis that seeks to assess the impact of political opportunity structures on political parties is altogether different, however. Here the aim is to investigate the opportunities and incentives that a given (subnational) context affords parties and politicians, and crucially, we would contend that turnout is not part of that context. Since the (local) turnout is clearly not known to anyone before the evening of the election day, we would argue it is neither part of an opportunity structure nor a general contextual variable that could somehow affect the probability of a radical right vote.

 

We certainly recognize that the level of turnout might reflect the attitudinal atmosphere or the intensity of political competition in a particular locality, and that this in turn, may indeed be important in explaining the success of a political party. However, we have concerns about using turnout as an (ex-post facto) indicator of attitudinal atmosphere or political competition because turnout will be affected by a whole host of other factors including the political tradition of an area, the specific local issues, the personalities involved in the campaign, and even the weather. As such, interpreting the cause of differing levels of turnout is highly problematic.

 

With their last two variables, Kestilä and Söderlund assess the effect of immigration and unemployment on the FN vote. Their model includes the share of the population of each départment that is made up of immigrants born outside of the EU-15 and of unemployed people.[7] This, however, is problematic because the coefficients for immigration and unemployment pick up at least three different things: i) they may represent a true contextual effect whereby immigration and unemployment provide the FN with an incentive to mobilize voters and whereby voters who feel strongly about these issues have an opportunity to vote for a party that campaigns on them; ii) they pick up the effect of the composition of the départements; and iii) they reflect cross-level interactions between features of the context and features of individuals.[8]

 

This becomes clear if we consider départements with a high share of immigrants. If we assume that the presence of immigrants facilitates mobilization by the FN, then people living in such départements should, ceteris paribus, have a higher propensity to vote for the radical right. This contextual effect should therefore result in a positive aggregate correlation that reflects a subnational political opportunity structure.

 

However, things are not that simple because we also need to bear in mind the composition of départements and their immigrant population. In 1999 (the year of the census on which Kestilä and Söderlund rely), there were 4.0 million people living in France who had been born outside the EU-15.[9] However, the majority of these immigrants (57 per cent) were French citizens and, as such, had the right to vote. Presumably, they and any of their children born in France, as well as children born in France to non-naturalized immigrants, and many of these people’s friends will have a probability of voting for the FN that is close to zero. Everything else being equal, therefore, these individual effects will result in a substantial negative aggregate correlation that counteracts the positive relationship resulting from the contextual effect.

 

Given that in roughly one third of all départements immigrants born outside of the EU make up more than five per cent of the population, and that in some départements of the Ile-de-France and the Provence-Alpes-Côte d’Azur regions they comprise up to 20 per cent of the population, the individual effect is not negligible, something which is reflected in the bad model fit for the banlieues of Paris (see below). Finally, such a scenario implies a cross-level interaction too, in that while the FN will be able to mobilize more voters because of high number of immigrants, it will only be able to do so among non-immigrants.

 

The same logic applies to the effect of unemployment, too. Observed aggregate correlations between unemployment and the FN vote are the result of a contextual effect (voters respond to regional unemployment levels), a compositional/individual effect (the unemployed are presumably more likely to vote for the FN), and possibly also a cross-level interaction effect: after all, it seems reasonable to assume that the strength of the individual effect of unemployment varies with the prevalence of unemployment in one’s environment.

 

The aggregate correlations Kestilä and Söderlund present therefore conflate three conceptually different effects, the nature and size of which are impossible to separate without micro-data.[10] In addition, because these coefficients reflect the highly aggregated net result of different processes, they hide any co-variation that is likely to exist both between and among individual and contextual variables.[11] For these reasons, the coefficients for unemployment and immigration in Kestilä and Söderlund’s model do not provide reliable information on the role of unemployment and immigration within a political opportunity structure.

 

Estimating and interpreting the coefficients of Kestilä and Söderlund’s model

 

As the discussion above has demonstrated, Kestilä and Söderlund’s model is problematic because all of the independent variables included in it raise theoretical, conceptual and/or methodological concerns. In addition to the problems that beset individual variables, the number of cases in Kestilä and Söderlund’s model is not very large (N=94) and the units (départements) vary enormously in terms of their population. The standard deviation for this variable is .48 million people, and the distribution is substantially right-skewed. The number of inhabitants for the ten smallest départements varies between 77,000 and 190,000, whereas each of the ten largest départements has a population of between 1.3 and 2.6 million people. The implication is that a lot of information on individual behaviour is lost, and that the behaviour of citizens in large départements will, ceteris paribus, have a smaller impact on the aggregate correlations.[12]

 

However, even leaving aside these concerns, the effects of the different independent variables are either difficult to interpret or trivial. As mentioned already, the effects of immigration and unemployment cannot be unambiguously interpreted because both variables aggregate the individual characteristics of the voters of the 94 départements of mainland France and, in the process, conflate contextual and individual effects and cross-level interactions. As for the other variables contained in the model, even though they capture features of the départements that exist independently of the individuals living in them, as we will demonstrate, their political impact is small – a fact that is not apparent in Kestilä and Söderlund’s reading of their findings. Moreover, as we will also show, the estimates in Kestilä and Söderlund’s model are highly sensitive to the selection of cases.

 

Kestilä and Söderlund interpret their results mainly with reference to the relative size of the t-values, and this is problematic for three reasons. Firstly, the jury is still out on the question of whether it makes sense to calculate (classical) standard errors for data that are a population rather than a sample (Berk & Freedman 2003). Secondly, if significance tests are to be carried out, the calculation of the standard errors should take into account the spatial correlations that exist between départements. Ignoring these dependences violates the standard assumption that disturbances are identical and independently distributed. And thirdly, and most importantly, the size of a t-value (i.e. statistical significance) is not a criterion for substantive relevance, and so it assists little to an interpretation of the effects of the variable.

 

For these reasons, rather than focussing on the t-values, we would suggest that the effects of the variables are best interpreted by examining the expected change in the FN’s vote share for a given change of the independent variables. At the same time as examining this, it is also important to consider the distribution of the independent variables if we are going to be able to say anything about political realities.

 

Kestilä and Söderlund do show us what the expected change in the FN’s vote share is for changes in the independent variables. Indeed, although they do not discuss these expected changes in the text, in Table 3 of their article we can see that a unit increase in the logged district magnitude would reduce support for the FN by 3.45 percentage points, whereas a unit increase in the effective number of parties would increase the FN’s share by 1.14 points. However, what these results do not do is take into account the distribution of the district magnitude and the effective number of party lists across départements and as such, they tell us little about the true impact of these variables on particular départements.

 

Let us first consider the logged district magnitude in 1998 and its distribution across départements. By identifying the second and the third quartile we can ascertain that in half the départements the total number of seats to be filled was between 10 and 22. And we can work out that by increasing the district magnitude from 10 to 22 while holding all other independent variables constant the expected support for the FN would be reduced by a mere 2.7 points.[13] This suggests that the effect of the district magnitude is fairly small across these départements. What is more, if we were to examine the middle 90 per cent of the distribution instead, we would find that the expected difference between the smallest district of eight seats and the biggest district of 31 seats would be 4.7 points. This is still not very large, and here we are considering the vast majority of cases. Thus, even though the effect of the logged district magnitude is statistically significant, it seems that it is only really relevant when we consider very small and very large départements.[14]

 

When we repeat this exercise for the effective number of party lists, we see that increasing the effective number of party lists from 2.6 (the second quartile) to 3.5 (the third) would increase support for the FN by just one percentage point. And if we consider the middle 90 per cent of the distribution, where the effective number of party lists ranges from 2.2 to 4.2, we see that a change from 2.2 to 4.2 would give rise to a 2.3 percentage point increase in the FN’s vote share.

 

As well as taking into account the distribution of the independent variables across départements we also need to bear in mind that their effect can be conditional on the levels of the other four independent variables in each département. This is the case for district magnitude: the bivariate correlation between district magnitude and FN support is essentially nil in Kestilä and Söderlund’s model, and only becomes negative once both turnout and unemployment are included in the model. Yet if we move away from the overall model and consider different subgroups of départements, we see that the relationship is actually positive for départements with below-average unemployment and turnout levels, whereas it is negative if either turnout or unemployment or both are above average. This rings alarm bells because there is no obvious theoretical reason for this finding. As such, and particularly because the number of units is low, it points to the possibility that the negative effect of district magnitude (as well as the positive effect of the number of party lists) may be spurious and driven by outlying and otherwise unusual observations.

 

To investigate this suspicion further, we calculated a number of diagnostics (studentized residuals, Cook’s distance, and leverage values) that can be used to identify problems with the model fit. We found that one département – Seine-Saint-Denis, which, with the neighbouring départements of the Hauts-de-Seine and the Val-de-Marne, forms the infamous banlieues of Paris – clearly stood out. Seine-Saint-Denis has the highest share of immigrants born outside the EU and the second-largest population of that group in absolute terms, and yet levels of FN support here are far below what Kestilä and Söderlund’s model predicts: while they predicted that the FN would poll 25.5 per cent in this département, the actual result in 2004 was 15.8 per cent. This just goes to show what happens when the contextual and compositional effects of the immigration rate are conflated. Furthermore, the impact of this département on the model is large as it is an influential data point in terms of the independent variables and is the largest negative outlier. If it is excluded from the estimation the coefficient for the immigration variable almost doubles and becomes statistically significant.

 

The largest positive outlier, by contrast, is the Vaucluse in the Provence-Alpes-Côte d’Azur region. This département had an average district magnitude in 1998 and slightly above-average figures for all other independent variables. While Kestilä and Söderlund’s model predicts a vote share of 18.7 per cent for the FN in the Vaucluse, the actual result was a staggering 28.5 per cent. The high support for the FN in this département reflects a political tradition that dates back to the 1980s. In the legislative elections of 1986, Jacques Bompard, a founding member of the FN, polled 18 per cent for the party in this département – one of the best results for the party in that election. Bompard (who left the party in 2005) was also instrumental in the FN’s successes at the local and the regional level, and in 1995 he became mayor of Orange (a historical town in the Vaucluse), being one of the first members of the FN to hold such an office.[15]

 

Since the Vaucluse does not have much leverage (as regards the independent variables, it is pretty average in almost every way), excluding it from the estimation does not greatly affect the coefficients. However, with just 94 départements, the joint leverage of a small group of three or four cases can easily be a problem (Fox 1997: 281). Indeed, it is possible to manipulate the coefficients considerably by excluding a tiny fraction of the départements. For instance, excluding not only the Vaucluse but also Paris (i.e. département 75) and the Territoire de Belfort in the Franche-Comté region reduces the absolute value of the coefficient for the logged district magnitude from -3.4 to -2.8. By contrast, excluding Seine-Saint-Denis and two rural départements with low unemployment and immigration rates – the Cantal in the Auvergne and the Haute-Vienne in the Limousin – increases the coefficient to -4.3. Similarly, excluding Seine-Saint-Denis together with the Haut-Rhin in Alsace (a FN stronghold) and the Lot-et-Garonne in Aquitaine halves the coefficient for the effective number of party lists.

 

The most striking effect is observed if we consider the share of immigrants born outside the EU. This coefficient is rather small (.15) and statistically insignificant in Kestilä and Söderlund’s model. Excluding the Vaucluse and two other départements where the FN is very successful – the Ain in the Rhône-Alpes region and the Alpes-Maritime in Provence-Alpes-Côte d’Azur – further reduces the effect of immigration to .03. However, excluding Paris, Seine-Saint-Denis and either of the other banlieues départements (Val-de-Marne or Hauts-de-Seine) almost triples the coefficient and turns immigration in a powerful (and statistically highly significant) predictor of FN success. Not only does this illustrate just how sensitive the estimates in Kestilä and Söderlund’s model are to the selection of cases, but it also highlights once again that the model conflates contextual (i.e. opportunity structure) and compositional (i.e. individual) effects.

 

Clearly, one could question this practice of excluding individual départements for diagnostic purposes given that Kestilä and Söderlund are examining the population of French départements rather than a sample. That said, doing this does enable us to assess just how accurate an instrument Kestilä and Söderlund’s model is for examining the impact of subnational political opportunity structures on the FN vote in the regional elections of 2004, and indeed for making generalizations beyond this particular electoral contest.

 

To further investigate our concerns about the spurious nature of the effects of the variables included in Kestilä and Söderlund’s model, we introduced an alternative predictor into their model: the vote won by Jean-Marie Le Pen in each département in the first round of the 2002 presidential election. The theoretical relevance of this variable in the context of the regional elections of 2004 is clearly only modest. That is, while we do expect Le Pen’s 2002 vote score to be a strong predictor of the FN’s vote in the 2004 regional elections because this would demonstrate that FN support at the departmental level is stable over time, the main purpose of introducing this additional variable into the model is to observe what happens to the effects of the other independent variables.

 

We chose this particular variable because it allows us to control for the fact that, over decades, the FN has been much more successful in some parts of France than in others (Bréchon & Mitra 1992), something which is due to the stabilizing effect of local party organizations (Lipset & Rokkan 1967: 53) and to the compositional effects and structural factors that benefit the party. Furthermore, the vote won by Le Pen in 2002 is an attractive measure of FN entrenchment because it cannot possibly have been affected by the district magnitude and the effective number of party lists in the regional elections in 1998 as the 2002 election was held under a completely different electoral system. As such, including this new variable in the regression should yield unbiased results for the relationship between FN support in 2004 and district magnitude/party system fragmentation in 1998, net of any other (stable) factors that are related to FN success at the departmental level.

 

 

[Table 1 about here]

 

 

Table 1 presents the coefficients of Kestilä and Söderlund’s original model as well as those for the augmented model (column 2). As we expected, the vote for Le Pen in the presidential election of 2002 turns out to be a strong predictor of FN success in the regional elections of 2004: each percentage point increase in support in 2002 translates into an increase of .98 percentage points in the party’s vote in 2004.

 

More important, however, is the fact that, once the lepeniste vote is controlled for, all other factors except unemployment are of very minor importance, with estimates that are very close to zero.[16] The lack of relevance of the five original independent variables is confirmed by a further model (column 3 of Table 1) in which all of the five original predictors are dropped and which shows FN support in the 2004 regional elections to be essentially identical to Le Pen’s vote in 2002 minus a constant of 2.7 percentage points.[17]

 

The discussion above suggests that, in the first instance, the lack of robustness of Kestilä and Söderlund’s model means it is unable to provide a compact description of the FN’s success at the departmental level in the 2004 regional elections. However, from both of the alternative models presented in Table 1 we have to conclude that the features of the subnational political opportunity structure included in Kestilä and Söderlund’s original model were largely irrelevant in explaining the FN’s vote in the 2004 regional elections anyway. As such, Kestilä and Söderlund’s model does not enable reliable inferences to be made about the impact of contextual factors on the radical right vote in Western Europe more generally, let alone allow for inferences that are more reliable than those made in the existing cross-national studies.

 

 

Towards an alternative model of FN success in the 2004 regional elections?

 

While we have demonstrated that Kestilä and Söderlund’s analysis suffers from a whole host of conceptual and methodological problems, we are still convinced that subnational political opportunity structures can, in principle, be very useful in accounting for the electoral success of radical right parties (and indeed any other type of party) provided this concept is operationalized in a more stringent way.

 

Given the data at hand, and especially given the lack of micro-level data on immigration and unemployment, the most obvious way the model may be improved is by replacing the ‘effective number of party lists’ variable with a variable that captures the ideological nature of party competition in the regional elections of 2004. As we argued earlier, the effective number of party lists reflects the accessibility of the regional party system, which is rather irrelevant in the case of the FN. Moreover, this variable has an element of tautology to it as it is not independent of previous levels of support for of the party. We therefore suggest replacing the effective number of party lists with two very simple variables: i) the presence of a second ‘extreme right’ list presented by the Mouvement National Républicain (MNR), and ii) the number of lists submitted by parties of the moderate right. Information pertaining to the lists presented in each region is readily available from the French government’s website (www.interieur.gouv.fr/).

 

We would expect the presence of an MNR list to reduce support for the FN, albeit only slightly. Given that the MNR broke away from the FN in January 1999 and is led by Le Pen’s former deputy, Bruno Mégret, one would expect many voters to see this party as a substitute for the FN.[18] As such, the presence of an MNR list should, ceteris paribus, reduce support for the FN because the political space available to the FN is more crowded. That said, we anticipate that the effect will only be modest because the MNR’s challenge to the FN effectively collapsed with the 2002 presidential election (Kuhn 2005: 102), when Mégret picked up only 2.3 per cent of the vote in the first round while Le Pen won 16.7 per cent and went on to contest the second round against the incumbent president, Chirac.

 

The number of mainstream competitors should also have a negative effect on the FN’s vote. Since this effect will not necessarily be linear, we will distinguish between three different scenarios: the presence of a single mainstream right list; the presence of two such lists; and the presence of three or more.

 

 

[Table 2 about here]

 

 

As a point of reference, column 1 of Table 2 shows the regression of the FN’s vote share in 2004 on the effective number of party lists in 1998 – i.e. the indicator favoured by Kestilä and Söderlund. The effect of this variable is slightly stronger in this bivariate model than it was in Kestilä and Söderlund’s complete model, but the very low R2 shows that it explains only a tiny fraction of the variation in the FN’s support. What is more, as is evident in column 2, the effect of the effective number of party lists disappears completely if we control for entrenched FN support by once again introducing our alternative predictor (the vote won by Le Pen in the first round of the presidential elections of 2002) into the model.

 

Column 3 of Table 2 presents the results of a model based on our alternative operationalization of party competition. It includes a dummy variable which takes a value of 1 in each département where the MNR presented a list, and dummy variables for the presence of two mainstream right lists, and three or more mainstream right party lists. This alternative model clearly fits the data much better than the effective number of party lists model. It explains a larger share of the variance in the FN’s vote, and the lower Bayesian Information Criterion (BIC) indicates that even though it includes more independent variables (and hence loses two degrees of freedom), this alternative model is preferable to the effective number of party lists one.

 

In this model the coefficients for competition from the moderate right have a straightforward interpretation and confirm our expectations: the FN’s vote in 2004 is reduced in départements where there were multiple moderate right lists. Compared to départements where the moderate right presented just one list, the FN vote is substantially (by over 6 percentage points) reduced where the party faced two mainstream right party lists. Where there were more than two mainstream right lists, the FN’s vote is reduced by over 3 percentage points.

 

Contrary to our initial expectation, however, we see that the presence of an MNR list in the 2004 elections does not reduce support for the FN. Rather, the presence of an MNR list has a substantial positive effect on the vote of the FN in 2004: after controlling for party competition from the mainstream right, the FN is on average 3.2 percentage points stronger in départements where the MNR fields candidates. We might explain this unexpected positive effect by pointing to the strategic choices made by the MNR’s leadership. While the FN presented candidates in all regions (and all départements), the MNR, stretched for money and staff, focussed its efforts on regions where the radical right had done well in the past – i.e. in areas where it might expect to do well. It contested 11 of the 14 regions where then FN had won above-average results in 1998 but chose to fight in only 2 of the 7 regions where the FN’s performance was below its average in 1998 (Cramér’s V=.49). The coefficient therefore picks up both the negative impact of competition from the MNR as well as the positive effect of previous FN support.

 

Our tentative explanation is confirmed by the findings in column 4 of Table 2: once we introduce the now familiar indicator for entrenched FN support, the effect of a competing MNR list becomes negative, as expected. Moreover, the effects of competition from the moderate right remain negative, though they are substantially reduced. This latter finding might again reflect strategic considerations of the FN’s competitors. After all, there are clear incentives for the moderate right to present a unified list in FN strongholds, something which is evidenced by a substantial correlation of r=0.42 between the FN support in the preceding regional election and the presence of a single mainstream right list.

 

The most important point about the model presented in column 4 of Table 2 is that the presence of an MNR list and the fragmentation of the mainstream right continue to have a theoretically meaningful effect even if previous FN support is controlled for. Moreover, the BIC indicates that this is an improvement over both the model that combined Kestilä and Söderlund’s effective number of party lists and the lepeniste vote (column 2) and the ‘pure’ model of entrenched FN support from Table 1. We take this as evidence that local/regional ideological competition matters and that it should be included in a subnational political opportunity structure model for radical right parties. The same cannot be said for the effective number of party lists.

 

As regards the other variables in Kestilä and Söderlund’s model, there is, unfortunately, no ‘easy fix’. We believe that some of them – namely turnout and district magnitude in 1998 – should not be included in the model at all because their conceptual status is dubious. Replacing district magnitude in 1998 with district magnitude in 2004 is also not an option because there is effectively no variation in this variable due to the legal thresholds in operation. And as for immigration and unemployment, although these are clearly part of the subnational political opportunity structure, in the absence of micro-level data it is simply not possible to investigate their impact and to untangle their contextual, compositional and cross-level effects.

 

Given this situation it appears that a major data collection effort is required if subnational political opportunity structures are to be operationalized rigorously and analysed fully and we would argue that such an endeavour should really go beyond merging survey data with subnational immigration and unemployment figures because Lubbers and Scheepers (2002) have already conducted an analysis of this kind for France. Rather, in an ideal world, a prospective project should collect data on variables that capture the theoretical concept of subnational political opportunity structures. This might include a content analysis of the local and regional media so as to capture its tenor (see Boomgaarden & Vliegenthart 2007 for a recent application to the national media in the Netherlands), an assessment of the organizational strength of local parties (see Pedahzur & Brichta 2002 on the institutionalization of the FN), and in-depth interviews with local political elites to probe their stances on radical right issues.

 

 

Conclusion

 

In their article, Kestilä and Söderlund highlight an important point, which although sometimes discussed in theoretical terms (e.g. Eatwell 2003), has largely been overlooked in empirical studies of the success of the radical right in Western Europe: local and regional contexts should not be ignored. Unfortunately, however, the importance of this message is somewhat obscured by the actual analysis that Kestilä and Söderlund carry out. For the reasons outlined above, we believe that there are difficulties with both Kestilä and Söderlund’s conceptualization of subnational political opportunity structures and their empirical findings.

 

A large data collection exercise focussing on factors that capture the concept of subnational political opportunity structures could potentially resolve many of the problems that Kestilä and Söderlund encountered in their study. Moreover, if time and resources were invested in any such future project, it would be all the more useful to analyse the relevant variables in a cross-national perspective. After all, authors such as Lubbers and Scheepers (2001, 2002) and Dülmer and Klein (2005) have already applied standard models of radical right voting to subnational units in individual countries.

 

That said, we fully concede that constructing cross-national models of radical right voting that contain rich information on very small subnational units (smaller even than French départements) would be a substantial accomplishment. Although collecting suitable data on one country is possible – as the British Election Study demonstrated more than ten years ago – gathering appropriate, comparable data across many countries would be a formidable feat. What is more, a cross-national study of subnational political opportunity structures would have to grapple with difficulties that are inherent to comparative analyses of this kind. That is, it would have to deal with the trade-off that exists between being able to draw conclusions that may be generalized beyond the cases in question and being able to gain an understanding of the intricacies of the particular contexts being examined. Indeed, some of the difficulties that Kestilä and Söderlund faced in their study reflect this very point: on the one hand their model is very sensitive to the selection of cases and hence does not allow for generalizable inferences to be made beyond the context of the 2004 French regional elections, but yet, on the other, it does contain rich information on the characteristics of the French regions and départements. This trade-off between generalizability and richness of data might raise questions over the very utility of any cross-national study of subnational political opportunity structures. Yet, if an appropriate balance can somehow be struck between these two concerns, we might learn a great deal more about the impact of local and regional contexts on the vote for radical right parties.

 


Notes


References

 

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Figure 1: The relationship between the effective number of party lists and the vote share of the FN in four French départements in the 1998 regional elections

 

 

Table 1: Alternative models of FN support in the French 2004 regional elections

 

 

(1)

(2)

(3)

 

Kestilä & Söderlund’s model

Kestilä & Söderlund’s model plus Le Pen vote

Le Pen vote only

District magnitude 1998 (ln)

-3.447***

-0.352

 

 

(0.855)

(0.475)

 

Effective number of lists 1998

1.137*

-0.0582

 

 

(0.484)

(0.256)

 

Turnout 2004 (per cent)

-0.736***

-0.107

 

 

(0.126)

(0.0748)

 

Immigrants born outside EU (per cent)

0.150

-0.0791

 

 

(0.120)

(0.0623)

 

Unemployment (per cent)

1.582***

0.432*

 

 

(0.336)

(0.185)

 

MNR running

 

 

 

 

 

 

 

Moderate right lists: 2

 

 

 

 

 

 

 

Moderate right lists: 3+

 

 

 

 

 

 

 

Vote for Le Pen 2002

 

0.979***

1.042***

 

 

(0.0612)

(0.0450)

Constant

55.94***

3.852

-2.656**

 

(9.195)

(5.687)

(0.787)

Adj. R2

0.436

0.855

0.852

Root MSE

4

2

2

BIC

534

410

394

d.f.

6

7

2

Log-Likelihood

-253

-189

-192

N

94

94

94

Standard errors in parentheses

* p < 0.05, **p < 0.01, ***p < 0.001

 


Table 2: The effect of the effective number of party lists in 1998 and ideological competition in 2004 on FN support in the French 2004 regional elections

 

 

(1)

(2)

(3)

(4)

 

Effective number of party lists

Effective number of party lists plus Le Pen vote

Ideological Competition

Ideological Competition plus Le Pen vote

Effective number of lists 1998

1.402*

-0.0370

 

 

 

(0.607)

(0.249)

 

 

Vote for Le Pen 2002

 

1.044***

 

1.125***

 

 

(0.0468)

 

(0.0502)

MNR running

 

 

3.181**

-2.167***

 

 

 

(1.076)

(0.483)

Moderate right lists: 2

 

 

-6.411***

-0.255

 

 

 

(1.423)

(0.620)

Moderate right lists: 3+

 

 

-3.206*

-1.112

 

 

 

(1.434)

(0.567)

Constant

10.59***

-2.570*

16.93***

-2.013

 

(1.974)

(0.979)

(1.474)

(1.022)

Adj. R2

0.044

0.850

0.204

0.879

Root MSE

5

2

4

2

BIC

569

399

559

386

d.f.

2

3

4

5

Log-Likelihood

-280

-192

-271

-182

N

94

94

94

94

Standard errors in parentheses

* p < 0.05, **p < 0.01, ***p < 0.001

 

 


[1] The choice between these two terms seems to be largely a matter of taste. To avoid unnecessary confusion, we follow Kestilä and Söderlund who chiefly use the adjective ‘radical’.

[2] Since the findings for the index of electoral success are largely comparable, our discussion will focus on the more straightforward measure of electoral support (i.e. vote share).

[3] In other words, in this simulation we assumed that voters would move between the FN list and mainstream right lists (i.e. RPR, UDF, and ‘divers droite’ lists) but not between left and right blocs. We further assumed that movements within the right bloc would not trigger movements between other parties. If we drop these assumptions and instead suppose that support for the FN comes from and goes to all other parties the results are almost identical. In both simulations, the upper threshold is the FN’s empirical vote share plus ten percentage points. The lower threshold is either the empirical vote share minus 20 percentage points or zero.

[4] The Pearson correlation for these four curves, which picks up the linear component, varies between .73 (Vaucluse) and .96 (Haute-Vienne and Indre).

[5] The share of the vote a party must win in order to gain parliamentary representation is determined either by the district magnitude or by the existence of a legal threshold if that legal threshold overrides the impact of district magnitude. To ascertain whether it is the district magnitude or the legal threshold which determines the vote a party needs for representation, or indeed to compare electoral systems with and without legal thresholds, we can make use of either Taagepera and Shugart’s ‘effective magnitude’ (1989: 135–141), or Lijphart’s ‘effective threshold’ (1994:182-183, note 29). Using the latter, the formula for which is 75/(M+1), we can see that if there had been no legal threshold in place in the regional elections of 1998, parties would have needed to win 18.75 per cent of the vote to gain representation in the district with the smallest magnitude (the Lozère which had a district magnitude of 3), whereas they would have needed only 1.03 per cent of the vote to win representation in the départment with the largest magnitude (i.e. the Nord which, as the most populous département, had a district magnitude of 72). In the Nord the legal threshold clearly overrides the effects of district magnitude. Indeed, the point at which the legal threshold starts overriding the effect of district magnitude is 14, since a district magnitude of 14 implies an effective threshold of 5 per cent.

[6] The electoral system used in 2004 included a number of legal thresholds. The law stipulated that in order for a party list to proceed from the first round of the election to the second it had to win at least ten per cent of the valid votes in the region. Lists that won five per cent in the region could also proceed if they fused with a list that had won ten pent of the valid votes in the region. The party list which won an absolute majority at the first round (if this occurred) or a plurality at the second round was given an automatic 25 per cent of the seats. The remaining seats were distributed proportionally among all party lists that had won at least five per cent of the votes in the region (Kuhn 2005; www.interieur.gouv.fr/). As it turned out, no party list won an absolute majority in the first round of the 2004 elections in any of the 22 regions of metropolitan France so all contests went to a second round. Had there been no legal thresholds in place in 2004 parties would have been able to win seats with very small percentages of the vote: district magnitudes of between 43 and 209 would infer effective thresholds of between 1.7 per cent and 0.36 per cent (see note 5). As such, the legal thresholds overrode the effects of district magnitude in all cases.

[7] Figures that pertain to the number of immigrants born outside of the EU do not capture the racial, ethnic, and/or religious characteristics of immigrants, something which is less than ideal in this instance given the FN’s appeals centre on notions of race, ethnicity and religion.  That said, data on the racial, ethnic and religious attributes of immigrants in France have not been collected.

[8] When both micro- and macro-data are available, separating these effects (by way of a multi-level model) is relatively straightforward. When only macro-level data is at hand (as in this instance), things are much more difficult, however. Indeed, the interpretation of pure macro-data leads almost inevitably to cross-level inferences which are highly problematic unless very specific assumptions hold (Achen & Shively 1995; Alker 1969; Robinson 1950). These assumptions include the need for extreme distributions (i.e. départements with almost no unemployment or immigration and départements with almost full or no unemployment or immigration) which would then enable the calculation of a range of individual correlations that are compatible with the observed aggregate correlations. Even then, however, one would need to be very cautious. What is more, things are even more complicated in this instance because, since Kestilä and Söderlund are interested in contextual effects, their study is not a straightforward ecological analysis of individual behaviour. Rather, their interest in contextual effects means that they imply a two-level model: conditions vary at the level of the département; these then affect whether party lists are presented at all, whether parties chose to present individual or joint lists, and just how much parties try to mobilize voters; and these two factors then affect the behaviour of individual voters.

[9] This does not include people born in the départements d’outre-mer and the territoires d’outre-mer (DOM-TOM), which are considered part of France for census purposes.

[10] See note 8.

[11] For instance, immigrants have a well above-average propensity of being unemployed, but individual unemployment status will in all likelihood have a different effect on the probability of an FN vote for immigrants and non-immigrants.

[12] The latter problem could be rectified by weighting the départements according to their population. In the event the coefficients do not actually change that much if départements are weighted by population, though the coefficient for immigration is effectively reduced to zero and the adjusted R2 decreases.

[13] ln(22)-ln(10)×3.45

[14] What is more, these calculations ignore the fact that the effects of district magnitude are effectively cancelled out in districts with a magnitude of 14 or more because of the existence of a five per cent legal threshold – see note 5. Had the effects of this legal threshold been taken into consideration, the effect of district magnitude would have been even smaller.

[15] Bompard was re-elected as mayor of Orange in 2001. Then, in September 2005 he resigned from the FN and joined the Mouvement pour la France (MPF) three months later. He was again re-elected as mayor in March 2008.

[16] The augmented model also allows much better predictions than the original one: the adjusted R2 almost doubles, while the mean squared error of the prediction is reduced by roughly 50 per cent. Given that just one additional parameter is estimated, the drop in the log-likelihood is massive. Accordingly, the Bayesian Information Criterion (BIC), which relates the improved fit of a more complex model to the ‘costs’ of adding parameters, drops substantially, indicating that the augmented model is indeed preferable to the original one.

[17] This simple model fits the data almost exactly as well as the augmented model, resulting in an even lower BIC.

[18] Mégret announced his retirement from politics in May 2008 and the following month it was decided at the MNR’s National Council that the party would henceforth be led by the 7-member Executive Bureau.

 


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