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

van Buuren, Stef and Karin (CGM) Oudshoorn. 1999. Flexible Multivariate Imputation by MICE. http://web.inter.nl.net/users/S.van.Buuren/mi/docs/rapport99054.pdf (07.10.2005).

van der Brug, Wouter and Meindert Fennema. 2003. Protest or Mainstream? How the European Anti-Immigrant Parties Developed into two Separate Groups by 1999. In: European Journal of Political Research, 42: 55–76.

van der Brug, Wouter and Meindert Fennema. 2007. Causes of Voting for the Radical Right. In: International Journal of Public Opinion Research, 19: 474–487.

Weakliem, David L. 2002. The Effects of Education on Political Opinions. An International Study. In: International Journal of Public Opinion Research, 14: 141–157.

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.