Christian Religiosity and Voting for West European Radical Right Parties

 

This is the author’s version of the work. Please cite as:

    Arzheimer, Kai and Elisabeth Carter. “Christian Religiosity and Voting for West European Radical Right Parties.” West European Politics 32.5 (2009): 985-1011. doi:10.1080/01402380903065058
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    This article examines the relationship between Christian religiosity and the support for radical right parties in Western Europe. Drawing on theories of electoral choice and on socio-psychological literature largely ignored by scholars of electoral behaviour, it suggests and tests a number of competing hypotheses. The findings demonstrate that while religiosity has few direct effects, and while religious people are neither more nor less hostile towards ethnic minorities and thereby neither more nor less prone to vote for a radical right party, they are not ‘available’ to these parties because they are still firmly attached to Christian Democratic or conservative parties. However, given increasing de-alignment, this ‘vaccine effect’ is likely to become weaker with time.

    @Article{arzheimer-carter-wep-2009,
    author = {Arzheimer, Kai and Carter, Elisabeth},
    doi = {10.1080/01402380903065058},
    journal = {West European Politics},
    number = 5,
    pages = {985--1011},
    keywords = {eurorex, cp},
    title = {Christian Religiosity and Voting for West European Radical Right Parties},
    volume = 32,
    url = {http://www.kai-arzheimer.com/christian-religiosity-radical-right-voting.pdf},
    html = {http://www.kai-arzheimer.com/christian-religions-radical-right-parties/},
    abstract = {This article examines the relationship between Christian religiosity and the support for radical right parties in Western Europe. Drawing on theories of electoral choice and on socio-psychological literature largely ignored by scholars of electoral behaviour, it suggests and tests a number of competing hypotheses. The findings demonstrate that while religiosity has few direct effects, and while religious people are neither more nor less hostile towards ethnic minorities and thereby neither more nor less prone to vote for a radical right party, they are not 'available' to these parties because they are still firmly attached to Christian Democratic or conservative parties. However, given increasing de-alignment, this 'vaccine effect' is likely to become weaker with time.},
    year = 2009
    }

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.

 

 

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