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Abstract

Radical left and right parties are increasingly successful—particularly among the less well-off. We assess the extent to which this negative effect of well-being on radical voting is moderated by contextual factors. Our study suggests that less well-off citizens vote for radical parties mainly under favorable aggregate-level circumstances. We distinguish two possible mechanisms underlying this effect—relative deprivation and risk aversion—and find support for relative deprivation only among radical right voters and for risk aversion for both types of radical voters, yet with predictable differences between the radical left and right supporter bases. Economic hardship leads to radical right voting when the socioeconomic circumstances are favorable and to radical left voting when net migration is modest. Our findings suggest a genuine paradox of radicalism: individual economic suffering might foster left and right radicalism, but mainly when that suffering takes place amid favorable conditions at the aggregate level.

Keywords

elections public opinion and voting behavior European politics

Radical parties and politicians are on the rise. In the watershed election of 2016 in the United States, for instance, Donald Trump’s radical right (RR)-wing discourse beguiled many dissatisfied Republicans, whereas radical left (RL)-wing Bernie Sanders struck a chord with similarly unhappy Democrats. This follows on a longer trend of rising radical parties on the other side of the Atlantic. On the right, Marine Le Pen’s Front National (FN) is increasingly successful in France, and Norbert Hofer of the Freiheitliche Partei Österreichs (FPÖ) almost won the Austrian Fall 2016 presidential elections. On the left, parties like Syriza in Greece and Podemos in Spain have become serious challengers to the political establishment. Although these radical left and right parties and politicians strongly differ from each other in their main ideology, they share the features of being eurosceptic (Hooghe, Marks, & Wilson, 2002) and populist (Rooduijn & Akkerman, 2017).

Many studies have shown that radical parties are particularly successful among those who experience individual economic difficulties. In particular, scholars have unearthed a strong negative relationship between someone’s economic well-being and his or her inclination to vote for a radical left party (see Gomez, Morales, & Ramiro, 2016; Ramiro, 2016) or radical right party (see Lubbers, Gijsberts, & Scheepers, 2002; Rydgren, 2013; Werts, Scheepers, & Lubbers, 2013). Our understanding of this relationship is thin, however. Most importantly, the relationship between well-being and radical voting can be expected to be moderated by national socioeconomic and sociocultural conditions, a possibility suggested by a recent study of how socioeconomic inequality moderates the link between income and radical right voting (Han, 2016). Yet we know only very little about how other contextual variables moderate this relationship,¹ and we know virtually nothing about such moderation for radical left voting.

In this article, we explore whether and how the tendency of one’s individual economic well-being to affect one’s vote for a radical party depends on country-wide contextual factors. We argue in particular that the relationship between well-being and radical voting is likely moderated by national socioeconomic and sociocultural conditions, such as the performance of the national economy, social policy protection, and levels of immigration. How this is so, however, is something that we consider to be a priori uncertain, as different conceptions of radicalism and voting suggest that aggregate socioeconomic and sociocultural conditions could moderate the negative relationship between individual well-being and radical voting in two starkly contrasting ways. On one hand, unfavorable conditions could deepen the tendency of individual hardship to spur radicalism: Those who are less well-off might be expected to vote for radical parties particularly if macrolevel economy-related conditions are unfavorable. This is our deepening hypothesis. On the other hand, aggregate context can be expected to have precisely the opposite moderating effect on the link between individual hardship and voting. More favorable macroeconomic conditions like low unemployment or substantial welfare state expenditures might actually heighten economically vulnerable voters’ sense of relative hardship or embolden such voters to gamble on unproven radical parties. According to our dampening hypothesis, hence, the negative effect of economic well-being on radical voting might also become dampened rather than deepened by unfavorable conditions, and might, under very harsh circumstances, even disappear.

We explore these possibilities through analysis of seven waves of European Social Survey (ESS) data (from 2002 to 2014), covering 21 countries, 27 radical right parties, and 22 radical left parties. Our analysis reveals support for the dampening hypotheses—yet with important differences between radical right and radical left voting. Someone’s perceived economic position is negatively related to radical right voting mainly when and where unemployment and inequality are low and social welfare expenditure and gross domestic product (GDP) per capita are high. These aggregate economic factors are not as significant, however, in moderating how individual hardship spurs support for radical left voting. The moderating role of immigration, however, appears to be stronger with respect to radical left voting: Individual economic hardship only leads to radical left voting if the immigration level is low, a discrepancy that might reflect the distinct issue ownership of the immigration issue among radical right but not radical left parties. These findings provide important evidence that there might well be a paradox of individual and aggregate well-being in the politics of radical voting: Individual hardship spurs radical left and right voting, but only when the aggregate conditions are favorable, rendering radical experimentalism safe for vulnerable citizens.

This study hereby provides important insights into the roots of populist radicalism of the left and right—where we can see how aggregate socioeconomic and sociocultural conditions strongly condition how individual economic hardship plays out for voting. Along the way, the study also provides new insights into the extent to which radical right and radical left voting differ from each other. Very few studies have compared those who vote for radical left parties with those voting radical right, and those studies that have done so have mainly emphasized the commonalities of these radical electorates (see Lubbers & Scheepers, 2007; Visser, Lubbers, Kraaykamp, & Jaspers, 2014). We show that these electorates behave very differently under different circumstances. This helps us understand radical voting in general—a phenomenon that has become increasingly important on both sides of the Atlantic.

Radical Parties and Radical Voting

To understand radical right and radical left electorates, one must first carefully delineate the radical party families. Radical right parties are first and foremost “nativist” parties: They argue, in Mudde’s useful wording, that “states should be inhabited exclusively by members of the native group (“the nation”) and that non-native elements (persons and ideas) are fundamentally threatening to the homogeneous nation-state” (Mudde, 2007, p. 19). Modern radical right parties also tend to be populist: They portray “the good people” as exploited, betrayed, neglected, or corrupted by “an evil elite” (Hawkins, 2010). Although radical right parties can differ from each other regarding their positions on ethical and socioeconomic issues, they constitute a quite homogeneous party family—or, at least, not more internally heterogeneous than, for instance, the conservative or liberal party families (Ennser, 2012). It has been argued and empirically established that radical right parties are disproportionately popular among voters with lower socioeconomic positions (see Arzheimer & Carter, 2006; Kitschelt, 1995; Lubbers et al., 2002; Rydgren, 2013; Werts et al., 2013).

Yet from these findings it does not follow that citizens from lower socioeconomic strata are automatically inclined to vote for the radical right. It might be that these citizens are just as likely to vote for radical left parties. Radical left parties are radical in that they reject the socioeconomic structure of capitalism. They are left because they conceive of socioeconomic inequality as one of the major political challenges of our time and therefore argue that alternative economic (more redistributive) models should be introduced that redress inequality (see Bale & Dunphy, 2011; Dunphy & Bale, 2011; March, 2011; March & Rommerskirchen, 2015). Voters who experience economic difficulties have good reasons to feel attracted to the parties expressing such radical left messages. Interestingly, however, only a handful of studies have assessed the relationship between voters’ socioeconomic positions and their inclination to support the radical left. These studies do suggest, however, plenty of similarities between left and right radicalism and their connections to socioeconomic position. Recent studies show that those with lower socioeconomic positions are more likely to vote for a radical left party (Bowyer & Vail, 2011; Gomez et al., 2016; Ramiro, 2016; Sperber, 2010).² In one essential respect, however, the socioeconomic background of radical left voters differs from radical right voters: Radical left voters tend to be higher instead of lower educated (Ramiro, 2016). This educational difference between left and right radicalism is most likely due to the more cosmopolitan, less nativist stances that education fosters and radical left parties accommodate (Hainmueller & Hiscox, 2007; Weakliem, 2002). Some recent studies assessing radical left and radical right voting simultaneously have corroborated these broad patterns (Lubbers & Scheepers, 2007; Visser et al., 2014; Rooduijn et al., 2017).

Voters with lower socioeconomic positions might be inclined to vote for radical parties for various reasons. First, they could support them because they agree with these parties’ main ideological messages. It has been shown that when it comes to radical right voting, the effect of someone’s socioeconomic position is strongly a function of, or mediated by, attitudes toward immigration (see Ivarsflaten, 2008; Kriesi, Grande, Lachat, Dolezal, & Bornschier, 2008; Kriesi et al., 2006; Van der Brug, Fennema, & Tillie, 2000, 2005). In particular, those who experience economic difficulties are likely to have anti-immigration attitudes and in turn, and therefore, vote for a radical right party. Indeed, anti-immigration attitudes appear to be more widespread among the unemployed and those with lower incomes and a lower education level (see Hainmueller & Hiscox, 2007; Schneider, 2008; Semyonov, Raijman, & Gorodzeisky, 2006). Similarly, studies of radical left voting have suggested different mediation effects—left-wing attitudes reduce the negative effects of socioeconomic position on support for radical left parties (Ramiro, 2016; Visser et al., 2014)—indicating that those who experience economic difficulties are likely to be in favor of welfare redistribution and are therefore prone to vote for the radical left. And it has long been shown that those with lower incomes and the unemployed are significantly more likely to be in favor of welfare redistribution (see Eger, 2010; Finseraas, 2009; Mau & Burkhardt, 2009).

It could also be the case that these voters support radical parties because they are discontented with politics (see Bergh, 2004). Some scholars have found that citizens with economic difficulties and lower incomes are more likely to be distrustful toward politics (Catterberg & Moreno, 2005; Hakhverdian & Mayne, 2012; Van der Meer, 2010). Because radical parties on both the left and right of the political spectrum express the populist message that the political elite does not listen to ordinary citizens anymore (Rooduijn & Akkerman, 2017), such citizens share their discontent with these radical parties and can therefore be expected to feel attracted to them. Recent scholarship, indeed, has shown that political discontent affects voting for both the radical right and radical left (Schumacher & Rooduijn, 2013) and that political discontent also, at least to a certain extent, mediates the effects of (perceived) socioeconomic position for radicalisms (see Werts et al., 2013 for the radical right and Ramiro, 2016 for the radical left).

In these substantive ideational logics for both radical left and right, the idea is that socioeconomically vulnerable citizens vote for radical parties because their socioeconomic position leads them to hold certain attitudes (see Zhirkov, 2014): attitudes toward immigration for radical right voters, attitudes toward welfare redistribution for radical left voters, and discontented attitudes toward politics for both radical left and right voters. In addition to these particular mediating attitudes, others might matter as well. For instance, recent research has shown that when it comes to both radical right and radical left voting, general left–right attitudes and attitudes toward European integration are important mediators (see Ivarsflaten, 2005; Ramiro, 2016; Van der Brug et al., 2000; Werts et al., 2013).

In addition to mediating attitudes, however, there are also nonideological reasons why economically vulnerable citizens vote for radical parties. Voters who experience economic difficulties might consider a vote for a nonmainstream party for more strategic reasons. Vulnerable voters may see radical parties as their best hope for a better future not (just) because they agree with the main ideological position of this party or with its antiestablishment position or because they want to express their feelings of discontent, but because a radical party harbors the opportunity to cast a vote for a challenger that is radically different from the mainstream parties. Because radical parties promise far-reaching changes, and such promises might well be highly attractive for those who suffer from economic hardship, the radical parties offer these voters a viable alternative to established parties. Existing studies provide some support for the existence of such nonideological support for radical parties: Although the effect of someone’s socioeconomic position on voting for a radical party is strongly reduced by including ideological variables, in many cases, the effects of socioeconomic variables do not fully disappear after controlling for the relevant attitudes (see Ramiro, 2016; Werts et al., 2013). It can therefore be expected that those who experience economic difficulties are more likely to vote for a radical right or a radical left party compared to a mainstream party even if we control for relevant attitudinal variables.

Unfavorable Contexts: Deepening or Dampening Radical Voting Among the Less Well-Off?

How an individual’s economic position affects his or her inclination to vote for a radical party, however, is not likely to be the same across different aggregate contexts within which people live and work. Existing scholarship focused on radicalism offers insights suggesting that the implications of individual economic experience for radicalism likely depend on the aggregate national socioeconomic and sociocultural setting. Based on existing theorizing on radicalism, we can anticipate two competing sets of hypotheses. On the one hand, unfavorable socioeconomic circumstances at the national level might well deepen the negative effect of individual well-being on voting for radical left and right parties. On the other hand, such unfavorable national-level conditions might (also) have the opposite effect, dampening the negative relationship between a person’s economic position and his or her voting behavior.

Various studies suggest that unfavorable socioeconomic and sociocultural contexts might well deepen the negative effect of economic well-being on radical right voting. Arzheimer (2009) has shown that radical right parties benefit from unemployment. Werts et al. (2013) and Jackman and Volpert (1996) have found similar effects. Golder (2003) and Boomgaarden and Vliegenthart (2007) have shown that unemployment rates matter especially when immigration levels are high. We might expect the negative effect of aggregate-level variables to hold in particular for those who experience economic difficulties. After all, the less well-off are more likely to feel threatened by high levels of unemployment or inequality than those with more secure economic positions. Indeed, Han (2016) has shown that when economic inequality increases, the poor become more likely to vote for radical right parties, whereas the rich become less likely to vote for these parties.

Similar developments might be expected to take place vis-à-vis radical left voting. Although we are not aware of research directly assessing the relationship between aggregate-level variables and radical left voting, several studies suggest that a poorly performing economy fuels support for radical left attitudes (Blekesaune, 2007). After all, a poor performance of the economy makes voters more aware of the risk of losing their jobs and also raises the concerns of those who are already unemployed. Blekesaune and Quadagno (2003) have shown that, indeed, high levels of unemployment increase support for welfare. Similarly, Dallinger (2010) has shown that a poorly performing economy, in terms of, for instance, GDP, leads to more support for redistribution. Finseraas (2009) found that next to unemployment rates and GDP, economic inequality is also related with support for welfare redistribution: In countries with more inequality, citizens tend to be more in favor of welfare support. It might therefore be expected that aggregate-level conditions that pose threats to the less well-off fuel not only radical right but also radical left voting.

Such logic leads to our deepening hypothesis—for both radical left and radical right voting. The general argument underlying this hypothesis is that individual economic hardship will lead to radical voting in ways compounded by macrolevel challenges. After all, if this is the case, those who are least well-off will become even more uncertain about their economic position and will therefore also be more likely to cast a vote for a challenger that is radically different. However, if the aggregate-level circumstances are favorable, those who experience economic hardship might well be likely to reward the political mainstream. Such a dynamic can be expected to hold also for the provisions of government most focused on redressing individual hardship: welfare state protection, such as actual spending on welfare transfers and services. It can be expected that the individual economic vulnerability of voters may spur support for radical left or right parties when such protection is low. Together, these intuitions support our general deepening hypothesis:

Hypothesis 1 (H1, deepening hypothesis): The negative effect of an individual’s well-being on his or her likelihood to vote for a radical party (either left or right) compared to a mainstream party will be more pronounced if the socioeconomic circumstances are unfavorable—that is., if the unemployment rate is high (H1a), if inequality levels are high (H1b), if GDP per capita is low (H1c), and if social welfare expenditure is low (H1d).

To some extent, this reasoning is consistent with the idea of economic voting: Citizens vote in favor of the government if the economy is performing well, but against the government if it is not (see Lewis-Beck & Stegmaier, 2000). Note, however, that our expectation does not concern government versus nongovernment parties but mainstream parties versus radical parties. This is an important difference, because mainstream parties can also be nongovernment parties and radical parties can also be government parties. Moreover, we do not focus on the effect of aggregate economic variables on voting in general but on the extent to which these variables moderate the relationship between individual well-being and radical versus mainstream voting.

Despite the intuitive appeal of our deepening hypothesis, the moderating role of aggregate economic context need not be only or mainly in the direction of deepening or compounding the effects of individual economic position. Indeed, there are good reasons to expect that external economic circumstances dampen rather than deepen the negative effect of economic well-being on radical voting. We distinguish two possible mechanisms underlying such a dampening effect: (a) relative deprivation and (b) risk aversion.

The literature on relative deprivation (see Crosby, 1976; Runciman, 1966; Walker & Pettigrew, 1984) predicts that in the context of a well-performing economy, those who are less well-off might well be more inclined to vote radical because they perceive themselves as being relatively even more deprived. According to Smith, Pettigrew, Pippin, and Bialosiewicz (2012) relative deprivation can be defined in terms of three steps. First, it can only be present if an individual compares himself or herself with other people, groups, or themselves at earlier points in time. Second, this comparison should lead to the perception that the individual or his or her in-group is at a disadvantage. Third, this perceived disadvantage should be conceived of as unfair. This leads to the feeling that one does not get what she or he deserves and that, in turn, causes angry resentment. This theory thus suggests that those who are less well-off but live under socioeconomically favorable circumstances might well feel more deprived than those who experience similar hardship but under socioeconomically less favorable circumstances. The reason is that they benchmark their own problematic economic circumstances against the favorable socioeconomic conditions at the national level. If this benchmarking results in feelings of unfairness—and that might well be the case—it could well fuel resentment toward mainstream political parties. And this might, in turn, spur both radical left and radical right voting. This leads to the following dampening hypothesis with respect to the same macrolevel parameters on which our deepening hypothesis focuses:

Hypothesis 2 (H2, relative deprivation dampening hypothesis): The negative effect of an individual’s well-being on his or her likelihood to vote for a radical party (either left or right) compared to a mainstream party will be less pronounced if the socioeconomic circumstances are unfavorable—that is, if the unemployment rate is high (H2a), if inequality levels are high (H2b), if GDP per capita is low (H2c), and if social welfare expenditure is low (H2d).

A second reason why unfavorable socioeconomic circumstances might dampen instead of deepen the negative effect of well-being on radical voting is that although voting for a radical party might be the best hope for someone who experiences economic difficulties and faces external economic threats, it is also a risky strategy. It is far from certain that radical parties will actually make things better. In fact, because they generally propose radical changes from the status quo, of which the (economic) effects are highly uncertain, and because they often lack office experience, it is possible that, if these parties reach office, instead of making things better for the economically vulnerable, they make things worse. This is not just a possibility that mobilizes ardent opponents of radical parties but might also inform that positions of those sympathetic to the radical outsider quality of radical left and right parties. Such, in any event, is our expectation.

It could therefore also be expected that those who individually experience difficulties of getting by will only vote for a radical party if they do not experience aggregate national external threats to their own economic position and consider such a radical choice relatively “safe.” According to this line of reasoning, it can be expected that if such individually vulnerable voters do experience external threats, they become more risk averse and refrain from voting for a radical challenger. In other words, it can be expected that under unfavorable conditions the negative effect of economic well-being on radical voting becomes less instead of more pronounced and might under economically very unfavorable circumstances even disappear.

It can be expected, however, that radical right voting is risky under different circumstances than radical left voting. Radical right voting is a risky strategy for those who experience economic hardship if the economic conditions in a country are unfavorable. After all, plenty of research suggests that radical right parties focus on sociocultural issues instead of socioeconomic ones. They are, in other words, the “issue owners” of sociocultural issues like—especially—immigration (see Boomgaarden & Vliegenthart, 2007; Walgrave & De Swert, 2004). Because these parties often do not have a very clear agenda on socioeconomic issues and, moreover, do not pay much attention to these themes, voters might well consider it doubtful if they are the right parties to do something about unfavorable socioeconomic conditions. This leads to the following expectation:

Hypothesis 3 (H3, risk aversion dampening hypothesis vis-à-vis radical right): The negative effect of an individual’s well-being on his or her likelihood to vote for a radical right party compared to a mainstream party will be less pronounced if the socio economic circumstances are unfavorable—that is, if the unemployment rate is high (H3a), if inequality levels are high (H3b), if GDP per capita is low (H3c), and if social welfare expenditure is low (H3d).

Radical left voting, in contrast, can be expected to have a sharply contrasting profile with respect to risk avoidance. Radical left voting is not a risky strategy for the less well-off under conditions of economic misfortune—for instance, if the levels of unemployment and inequality are high and GDP per capita and welfare spending are low. After all, radical left parties strongly focus on economic issues and promise the radical redistribution of incomes. This may confer an element of issue ownership on distributional losses during economic bad times. However, radical left voting might well be considered a risky strategy for the less well-off when the level of immigration in a country is high. With high levels of immigration, those who experience economic difficulties might well fear that immigrants steal their jobs. Under such circumstances, voting for a radical left party might be considered risky because these parties strongly focus on socioeconomic issues and do not pay much attention to sociocultural issues like immigration. Moreover, if these parties pay attention to such issues, they tend to hold rather “multicultural” positions (March, 2011). It is therefore highly unlikely that, if in power, radical left parties would try to curb immigration. Radical right voting, however, is not a risky strategy if the level of immigration is high. On the contrary, if the level of immigration is high, radical right voting could be considered a rather “safe” strategy, because radical right parties strongly emphasize the immigration issue and, moreover, hold radical restrictive positions vis-à-vis immigration. This leads to the following expectation:

Hypothesis 4 (H4, risk aversion dampening hypothesis vis-à-vis radical left): The negative effect of an individual’s well-being on his or her likelihood to vote for a radical left party compared to a mainstream party will be less pronounced if the socio cultural circumstances are unfavorable—that is, if immigration levels are high.

Hence, regarding radical right voting, the risk aversion mechanism predicts similar outcomes as the relative deprivation mechanism. But vis-à-vis radical left voting, the two mechanisms lead to different expectations. In general, our dampening hypotheses represent an interesting paradox: They suggest that those with unfavorable economic positions are more likely to vote for a radical party only if the circumstances are favorable. An explanation for such a “paradox of well-being” might be that at the individual level, an unfavorable economic position refers to actual economic hardship, while at the aggregate level, unfavorable conditions do not imply actual individual economic hardship but only the increased probability or the threat of such hardship—especially for those who are already less well-off. In other words, unfavorable conditions at the aggregate level have other implications than hardship at the individual level.

We have, in sum, two competing sets of expectations for how unfavorable contexts can be expected to either deepen (H1) or dampen (H2-H4) the tendency of an individual’s economic hardship to spur his or her likelihood to vote for radicalisms of the left or right. Both sets of expectations are, in our judgment and in the existing state of empirical and theoretical scholarship, equally plausible a priori. We believe, however, that empirical exploration of public opinion can shed considerable light on these possibilities to adjudicate whether socioeconomic and sociocultural contexts deepen or dampen how individual hardship spurs radicalism.

Data and Measurement

To test these hypotheses, we use ESS data (seven waves, every 2 years between 2002 and 2014).³ We selected those individuals who voted for a radical party or for one of the mainstream parties (i.e., a liberal, sociodemocratic, conservative, or Christian democratic party). Our categorization of radical right and radical left parties is largely based on Mudde (2007) and March (2011). For an overview of the selected radical right and radical left parties, see Table 1. Our dependent variables are based on the vote choice of our selected respondents. Respondents were asked which party they voted for during the last national general elections. Based on this question, we constructed two dichotomous dependent variables: radical right voting (0 = mainstream party and 1 = radical right party) and radical left voting (0 = mainstream party and 1 = radical left party).⁴

Table 1.

Selected Countries and Parties.

Country	Radical right party	Radical left party
Austria	FPÖ, BZÖ
Belgium	VB, FNb
Czech Republic		KSCM
Denmark	DF	EL, SF
Finland	PS	VAS
France	FN, MNR	PCF, LO, LCR
Germany	Republikaner, NPD	Linke
Greece	LAOS	KKE, SYN
Hungary	Jobbik, MIÉP	MP
Ireland		SF
Italy	LN, AN	PRC, Comunisti
Netherlands	LPF, PVV	SP
Norway	FrP	Rodt, SV
Poland	LPR, KNP
Portugal		PCP, BE
Slovenia	SNS, LIPA
Slovakia	SNS	KSS
Spain		IU
Sweden	SD	V
Switzerland	SVP
United Kingdom	BNP, UKIP

Our main individual-level independent variable is economic well-being. It measures the extent to which individuals experience economic difficulties by asking respondents to report how they feel about their household income on a scale ranging from 1 (finding it very difficult on present income) to 4 (living comfortably on present income). In our baseline analyses, we do not look at actual income as a measure of economic well-being for two main reasons. First, we are interested in someone’s substantive interpretation of his or her economic situation. It might well be the case that person A experiences a certain income as low, while person B can cope rather well with this same amount of money. We are mainly interested in how people feel about their financial situation. Second, a more practical reason for not looking at actual income is that this variable has been measured much less consistently in the ESS. Including it significantly reduces the sample size. We control for various sociodemographic variables. We measure a respondent’s level of education by means of an assessment of the highest completed level of education. We distinguish it in five categories: (a) less than lower secondary education, (b) lower secondary education completed, (c) upper secondary education completed, (d) postsecondary nontertiary education completed, and (e) tertiary education completed. We also assessed whether respondents are unemployed (1 = unemployed), their age, their gender (1 = female), their subjective religiosity (11-point scale: 0 = not at all religious, 10 = very religious), and the rural/urban origin of the respondent (rural = 0, urban = 1).⁵

We also include as individual-level controls a set of attitudinal variables. We measure a person’s general left/right position with a scale that ranges from 0 (left) to 10 (right). We assessed the anti-immigration attitude of respondents by constructing a scale that consists of three variables that were measured by the following questions: (a) “Would you say it is generally bad or good for [country]’s economy that people come to live here from other countries?”; (2) “Would you say that [country]’s cultural life is generally undermined or enriched by people coming to live here from other countries?”; and (3) “Is [country] made a worse or a better place to live by people coming to live here from other countries?” The new variable was recoded so that 0 indicates that immigrants are good rather than posing a threat and 10 means that immigrants pose a large threat (Cronbach’s alpha = .84). We assessed support for redistribution by means of the degree to which respondents agree with the following claim: “The government should take measures to reduce differences in income levels.” The answering categories range from 1 (strongly disagree) to 5 (strongly agree). On an 11-point scale, respondents could indicate to what extent they trust the country’s parliament and politicians. We recoded these items so that they range from 0 (no trust at all) to 10 (complete trust) and combined them into a political trust scale (Cronbach’s alpha = .83). We control for satisfaction with the government and satisfaction with the present state of the economy, both on a scale from 0 (extremely dissatisfied) to 10 (extremely satisfied).⁶

We include various aggregate-level variables that we consider important as measures of economic context that we expect to positively or negatively moderate the effects of individual socioeconomic position. We measure the unemployment rate by assessing the percentage of unemployed people (compared to the total labor force; Organisation for Economic Co-operation and Development). We also include GDP per capita in our models. Socioeconomic inequality is measured by the Gini coefficient. We used data from the Standardized World Income Inequality Database (SWIID; Solt, 2014). To measure the size of the welfare state, we look at social welfare expenditure (in percentage of GDP). All these variables are lagged 1 year, meaning that a respondent’s sociodemographics and positions in 1 year are linked to the aggregate-level variables in the previous year. Net migration is measured as a percentage of the population (World Development Indicators [WDI], 2016). The values we have included are 5-year estimates (in 1997, 2002, 2007, and 2012), so we linked these estimates to the individual values measured in the next ESS wave. See Table A1 of Online Appendix for an overview of all included variables.

In our baseline models, we estimate logistic multilevel random intercept models in which individuals (Level 1) are nested in country-years (Level 2). The reported coefficients are odds ratios, and standard errors are robust-clustered by country-year. In the models in which radical right voting is the dependent variable, we included only country-years in which one of the radical right parties we have selected participated (n at Level 1 = 70,816, n at Level 2 = 82), and in the models in which radical left voting is the dependent variable, we included only country-years where one of the radical left parties we have selected participated (n at Level 1 = 76,661, n at Level 2 = 81). We also consider a range of alternative specifications, including alternative embedding of the multilevel models, alternative measures of radicalism, alternative measures of macrolevel economic and sociocultural conditions, and alternative estimators.

It is important to emphasize that we do not test the logics of relative deprivation and risk aversion directly, because our data simply do not allow for such tests. However, our assessment of the extent to which aggregate-level variables moderate the negative effect of well-being on radical right and left voting might still legitimate inferences about which of the two mechanisms is more likely (see the concluding section).

Findings

Table 2 reports the results of the multilevel regression analyses in which our dependent variable is radical right voting. In Model M₁, we have included all individual-level sociodemographic variables. Economic well-being has a significant negative effect, confirming the general expectation that those who experience economic difficulties are more likely to vote for the radical right. Also, those who completed tertiary education are less likely to vote for the radical right. Furthermore, supporters of the radical right are more common among younger, male, nonreligious individuals coming from rural areas.

Table 2.

Multilevel Logistic Regression Models Estimating Radical Right Voting (Odds Ratios Reported).

	M1	M2	M3	M4	M5	M6	M7	M8
Sociodemographic variables
Economic well-being	0.812***	0.933*	0.931*	0.822***	1.297**	0.514*	1.621**	0.891*
Education (reference = less than lower secondary)
Lower secondary	1.197*	1.314**	1.313**	1.310**	1.314**	1.301**	1.324**	1.314**
Upper secondary	0.937	1.175	1.174	1.171	1.168	1.162	1.180	1.176
Postsecondary	0.821	1.095	1.087	1.089	1.088	1.078	1.091	1.088
Tertiary	0.329***	0.561***	0.559***	0.558***	0.558***	0.554***	0.563***	0.560***
Unemployed	1.194	1.183	1.181	1.194	1.198	1.183	1.179	1.177
Age	0.989***	0.986***	0.986***	0.986***	0.986***	0.986***	0.986***	0.986***
Gender	0.651***	0.644***	0.643***	0.642***	0.642***	0.643***	0.643***	0.643***
Religiosity	0.943***	0.948***	0.950***	0.950***	0.949***	0.950***	0.950***	0.950***
Rural/urban	0.888**	0.990	0.991	0.990	0.988	0.991	0.988	0.991
Attitudinal variables
Left/right		1.288***	1.288***	1.288***	1.289***	1.288***	1.288***	1.288***
Anti-immigration		1.373***	1.372***	1.373***	1.372***	1.372***	1.372***	1.372***
Support for redistribution		1.082***	1.082***	1.082***	1.081***	1.082***	1.082***	1.082***
Political trust		0.840***	0.840***	0.840***	0.840***	0.840***	0.840***	0.840***
Satisfaction with the government		0.938**	0.938**	0.938**	0.938**	0.938**	0.939**	0.938**
Satisfaction with the present state of the economy		0.994	0.992	0.992	0.992	0.993	0.992	0.992
Aggregate-level variables
Unemployment rate			1.060	1.007	1.068	1.061	1.064	1.058
GDP			1.000***	1.000***	1.000***	1.000***	1.000***	1.000***
Gini			0.906	0.905	0.907	0.847**	0.906	0.906
Social welfare expenditure			0.984	0.980	0.978	0.983	1.054	0.985
Net migration			1.199	1.199	1.209	1.200	1.195	1.100
Interactions
Unemployment rate × Economic well-being				1.019*
GDP × Economic well-being					0.999***
Gini × Economic well-being						1.023*
Social welfare expenditure × Economic well-being							0.978**
Net migration × Economic well-being								1.027
Constant	0.536**	0.021***	0.026*	0.041*	0.010**	0.165	0.005**	0.030*
Country-year variance	3.408***	6.063***	2.840***	2.864***	2.811***	2.806***	2.838***	2.852***
n	70,816	70,816	70,816	70,816	70,816	70,816	70,816	70,816

GDP = gross domestic product.

p < .05. **p < .01. ***p < .001.

In Model M₂, we have added the individual-level attitudinal variables. It turns out that those who are more right-wing, more anti-immigrant, more in favor of redistribution, more distrustful toward politics, and less satisfied with the government are more likely to vote for the radical right. Including these attitudinal variables reduces the negative effects of economic well-being and education but does not render the regression coefficients of these variables statistically insignificant. This indicates that the effect of economic well-being is, as expected, not fully mediated by these attitudinal variables. The effect of rural/urban, however, does not exert a significant effect anymore after including these attitudes. The effects of age, gender, and religiosity remain largely unaltered after including these attitudes.

Model M₃ tests what happens if we also include our aggregate-level variables. The only aggregate-level variable that turns out to significantly affect radical right voting is GDP per capita. Including the aggregate-level variables does not alter the effects of the individual-level variables. The cross-level interaction effects of each of the aggregate-level socioeconomic variables with economic well-being are included in the Models M₄ to M₇. The interaction of economic well-being with the unemployment rate, GDP per capita, the Gini index, and social welfare expenditure is statistically significant (see Models M₄-M₇).

The marginal effects of economic well-being conditional upon the unemployment rate, GDP per capita and the Gini coefficient, welfare expenditure, and net migration are displayed in Figure 1. The first panel shows the effect conditional on the unemployment rate. It demonstrates that the effect of well-being is only significantly negative when the unemployment rate is rather low; when it exceeds about 7% (45th percentile), the effect is not significant anymore. The moderation effect of GDP per capita is shown in the second panel. It shows that the effect of well-being is significant only when GDP per capita exceeds 35,000 (50th percentile). The third panel shows the effect of well-being for the different values of the Gini coefficient. The effect is only significant for lower Gini values: from a Gini coefficient of about 27 onward (60th percentile) the effect is not statistically significant anymore. The fourth panel concerns social welfare expenditure. Here, we see that the effect of economic well-being is only significant for higher values: only from a value of 24% onwards (45th percentile) is the effect negative and significant. Interestingly, the effect of well-being is even positive for a very low value of social welfare expenditure. We need to be very careful, however, with our interpretation of this finding: As the histogram shows, the number of cases with such a low level of expenditure is very low. These findings are in line with H2 and H3: Those who experience economic difficulties are only more likely to vote for a radical right party compared to a mainstream party if the socioeconomic circumstances are favorable (i.e., if the unemployment rate is low, if GDP per capita is high, and if inequality levels are low) and if social protection is high (i.e., social welfare expenditure is substantial).

Figure 1.

Marginal effects of economic well-being on radical right voting conditional upon the unemployment rate, GDP per capita, the Gini coefficient, and social welfare expenditure.

In Table 3 we assess the effects on radical left voting. Again, Model M₁ displays the effects of the sociodemographic variables only. Economic well-being again exerts a negative effect: Those who have more difficulties of getting by are more likely to vote for the radical left. Also, the unemployed are more likely to vote for the radical left. As expected, the effect of education differs between radical left and radical right voters: Radical right supporters tend to be lower educated, whereas radical left voters are more likely to be higher educated. Furthermore, the radical left turns out to be popular among nonreligious females living in urban areas.

Table 3.

Multilevel Logistic Regression Models Estimating Radical Left Voting (Odds Ratios Reported).

	M1	M2	M3	M4	M5	M6	M7	M8
Sociodemographic variables
Economic well-being	0.690***	0.841***	0.840***	0.801***	0.909	0.808	0.787	0.782***
Education (reference = less than lower secondary)
Lower secondary	1.280*	1.247	1.238	1.235	1.233	1.236	1.239	1.232
Upper secondary	1.188	1.169	1.159	1.157	1.155	1.157	1.161	1.152
Postsecondary	1.170	1.166	1.155	1.153	1.151	1.153	1.157	1.150
Tertiary	1.374*	1.371**	1.359*	1.357*	1.354*	1.356*	1.360*	1.351*
Unemployed	1.337***	1.196	1.197	1.204	1.198	1.197	1.197	1.201
Age	0.998	0.993**	0.993**	0.993**	0.993**	0.993**	0.993**	0.993**
Gender	1.170***	1.013	1.012	1.012	1.012	1.012	1.012	1.010
Religiosity	0.836***	0.905***	0.905***	0.905***	0.905***	0.905***	0.905***	0.905***
Rural/urban	1.154***	1.103*	1.104*	1.103*	1.103*	1.104*	1.104*	1.104*
Attitudinal variables
Left/right		0.578***	0.578***	0.578***	0.578***	0.578***	0.578***	0.578***
Anti-immigration		0.943***	0.943***	0.944***	0.943***	0.943***	0.943***	0.943***
Support for redistribution		1.306***	1.307***	1.306***	1.307***	1.307***	1.307***	1.305***
Political trust		0.943***	0.942***	0.943***	0.943***	0.942***	0.942***	0.943***
Satisfaction with the government		0.894***	0.895***	0.894***	0.895***	0.895***	0.895***	0.895***
Satisfaction with the present state of the economy		1.003	1.001	1.001	1.001	1.001	1.001	1.001
Aggregate-level variables
Unemployment rate			0.934**	0.917*	0.934**	0.933**	0.933**	0.933**
GDP			1.000**	1.000**	1.000**	1.000**	1.000**	1.000**
Gini			1.016	1.015	1.016	1.012	1.016	1.016
Social welfare expenditure			0.991	0.990	0.990	0.991	0.983	0.993
Net migration			0.888***	0.888***	0.888***	0.888***	0.888***	0.776***
Interactions
Unemployment rate × Economic well-being				1.006
GDP × Economic well-being					1.000
Gini × Economic well-being						1.001
Social welfare expenditure × Economic well-being							1.003
Net migration × Economic well-being								1.045**
Constant	0.435***	2.221**	1.115	1.322	0.917	1.249	1.339	1.337
Country-year variance	1.491**	2.117***	1.667***	1.668***	1.661***	1.668***	1.666***	1.681***
n	76,661	76,661	76,661	76,661	76,661	76,661	76,661	76,661

GDP = gross domestic product.

p < .05. **p < .01. ***p < .001.

The attitudinal variables are included again in Model M₂. It turns out that those who are left-wing, proimmigration, proredistribution, politically distrustful, and dissatisfied with the government are more likely to vote for the radical left. After including these attitudinal variables, the effects of economic well-being decreased but, again, did not become insignificant. The effect of education remained positive. The effects of unemployment and gender disappeared after including the attitudinal variables.

Including the aggregate-level socioeconomic variables did not change the effects of the sociodemographic and attitudinal variables (see Model M₃). Of the aggregate-level variables, the unemployment rate, GDP per capita, and net migration exert significant effects on radical left voting. Interestingly, the effect of unemployment is negative, indicating that people are more inclined to vote for the radical left if the unemployment rate is low. The interaction effects of the aggregate-level socioeconomic variables are again included in the Models M₄ to M₇. The marginal effects are displayed in Figure 2. The first panel shows the effect of well-being conditional on the unemployment rate. The effect is negative and significant until the unemployment rate reaches a value of about 14%. Yet only about 5% of the cases surpass this value, so we can conclude that the effect of well-being on radical left voting is not conditional upon the unemployment rate. The Panels 2, 3, and 4 show that the effect of well-being on radical left voting is negative no matter what values of GDP per capita, the Gini coefficient, or social welfare expenditure. These findings for radical left voting go against not only the “deepening” (H1) but also our “dampening” hypothesis focused on the mechanism of relative deprivation (H2). Such a pattern suggests that an economically vulnerable voter’s attraction to radical left parties is less reflective of actual resentment or relative deprivation (manifest in one’s suffering in a national context of prosperity) than applies to attraction to the radical right.

Figure 2.

Marginal effects of economic well-being on radical left voting conditional upon the unemployment rate, GDP per capita, the Gini coefficient, and social welfare expenditure.

To see if the voting patterns are consistent with the risk aversion dampening mechanism, we must also look at the interaction of well-being with net migration. The results of these interaction effects are presented in the Models M₈ in Table 2 (radical right voting) and Table 3 (radical left voting). Interestingly, the interaction effect is insignificant vis-à-vis radical right voting, whereas it is statistically significant regarding radical left voting. The marginal effects are displayed in Figure 3. The panel on the left shows that, indeed, the cross-level interaction between economic well-being and net migration is not statistically significant regarding radical right voting. The negative effect is statistically significant only for net migration levels between 0 and 2. The panel on the right displays the significant interaction between well-being and net migration vis-à-vis radical left voting. For values of net migration lower than about 4% (85th percentile), the regression coefficient is significantly negative. For higher values of net migration, however, the effect is not significant anymore. Such patterns do corroborate the risk aversion hypotheses regarding radical right voting (H3) and radical left voting (H4): Those with economic difficulties are generally more likely to vote for radical right parties whatever the state of net immigration but are only more likely to vote for a radical left party compared to a mainstream party if the level of immigration is low.

Figure 3.

Marginal effects of economic well-being on radical right voting (left panel) and radical left voting (right panel) conditional upon net migration.

Robustness Checks and Extensions

To assess the robustness of our findings, we have executed a series of robustness checks. The detailed results of these checks can be found in the Online Appendix. Here, we will present a brief overview of the main outcomes.

First, we have assessed whether it makes a difference if we use a slightly different version of our dependent variables: comparing radical right and radical left voting to voting for all other parties instead of mainstream parties only. Regarding radical right voting we find one small difference with the main analysis: The interaction between the unemployment rate and economic well-being is not statistically significant anymore. Yet because the effect is still in the same direction (and very close to statistical significance), and the interaction effects of economic well-being with the Gini coefficient and social welfare expenditure remain statistically significant, we conclude that the patterns we found in the main analysis generally still hold. We found no big differences vis-à-vis voting for radical left parties.

Second, it has been argued that the radical left party family is rather heterogeneous (Gomez et al., 2016; March, 2011)—much more so than the radical right (see Ennser, 2012). We therefore considered whether the results in our main analysis regarding voting for radical left parties apply to two subfamilies of the radical left: traditional radical left parties relative to mainstream parties on the one hand, and progressive radical left parties relative to mainstream parties on the other hand (see Online Appendix for more information about this categorization). We discovered that the interaction between net migration and economic well-being is not significant when it comes to voting for traditional radical left parties—although the effect is still in the same direction. We expect the reason to be that traditional radical left parties hold less multicultural positions vis-à-vis the immigration issue than progressive ones (March, 2011). An important consequence might be that someone who is less well-off and faces high levels of immigration in his or her country considers a vote for a traditional radical left party (which is not multiculturalist) less risky than a vote for a more progressive radical left party with multicultural stances. In general, this means that we can conclude that the interaction between net migration and economic well-being only holds for more progressive radical left parties.

Third, we consider various specifications of our main independent variable focused on economic hardship. Our main independent variable is individual economic well-being, which we measured on an ordinal scale with four categories (asking how respondents feel about their household income on a scale with the following categories: 1 = finding it very difficult on present income, 2 = finding it difficult on present income, 3 = coping on present income, and 4 = living comfortably on present income). In our main analysis, we have modeled this item as a continuous variable. Because this might affect our results, we have also assessed the effects of this item as an ordinal variable with three categories (we collapsed the first two categories because the first category contains less than 5% of the respondents in the sample) and a dichotomous variable (combining the first two and the second two categories). This does not lead to substantially different results.

We also consider the possibility that the patterns we explore might extend to an individual’s judgment of aggregate economic hardship. Although the ESS data gives few footholds for such exploration, we do consider the possibility that “satisfaction with the economy” might have implications for RR and RL moderated by aggregate socioeconomic conditions in line with our baseline examination focused on economic well-being. The baseline models above have revealed “satisfaction with the economy” to be in general insignificant in shaping RR and RL. But supplemental analysis suggests that such satisfaction operates very much in line with the baseline results for individual economic well-being—but only for radical right voting. Echoing the baseline specifications in Table 1, we see that satisfaction with the aggregate economy tends to diminish support for radical right parties but that this effect is dampened by unfavorable macroeconomic conditions (high unemployment and inequality, and low GDP per capita). The supplemental analysis more weakly echoes the patterns in Table 2 for radical left parties (e.g., net migration tends not to dampen how satisfaction with the economy lowers chances of radical left voting). Such results suggest that H2 and H3 may extend to how sociotropic economic concerns affect radical right voting but that H2 and H4 likely do not extend to how sociotropic concerns affect radical left voting.

Fourth, two variables which might exert an effect on both radical right and radical left voting—class and support for the European Union (EU)—have not been included in the main analysis, because that leads to a strong reduction in sample size. As a robustness check, we have assessed if including these variables changes the results (see Online Appendix for how these variables are measured). If we control for class, the interaction effect between net migration and economic well-being is now statistically significant. This is consistent with our expectations and is only a minor difference with the main analysis, where the effect is in the same direction. There are no differences regarding radical left voting. Regarding support for the EU, when we include this variable, the effect of the interaction between the Gini coefficient and well-being is not statistically significant anymore. However, because the effect is still in the same direction and the interactions with unemployment and social welfare expenditure remain unaltered, we conclude that this alternative result does not challenge our main conclusions.

Fifth, we are interested in the effects of variables that are measured at the individual level and the country-year level. Therefore, we estimated multilevel regression models in which individuals (Level 1) are nested in country-years (Level 2). We also considered alternative embedding of the multilevel models, such as taking countries as Level 2, yielding very similar results to the reported baseline models. It might also be the case, however, that voting for radical parties is affected by country-level or year-level (i.e., wave-level) variables. To take such possibilities into account, we have also included country-fixed and year-fixed effects in our models. This does not change the results.

Sixth, we consider different combinations of and new measures of macrolevel conditions relevant to economic hardship. Our baseline results are clearly corroborated in models running no or fewer of the other macrolevel (Level 2) parameters, potentially relevant to overfitting (Stegmueller, 2013; Van der Meer, te Grotenhuis, & Pelzer, 2010). Also, other economic macro variables than the unemployment rate, GDP, the Gini coefficient, and social welfare expenditure may exert an effect on radical right and left voting as well. We have therefore estimated the effects of additional variables—GDP growth and the so-called “at-risk-of-poverty rate”—as well as the interactions of these variables with economic well-being. The results corroborate our main findings. When it comes to radical right voting, the negative effect of an individual’s well-being on his or her likelihood to vote for a radical right party compared to a mainstream party is less pronounced if the economic circumstances are unfavorable (this also holds true for high at-risk-of-poverty rate). And, as in the main analysis, the relationship between economic well-being and radical left voting is not conditional on GDP (growth) and the at-risk-of-poverty rate.

Seventh, macrolevel socioeconomic conditions may affect radical vote choice not in terms of developments with respect to a given level of macroeconomic conditions obtaining in a country in a given year but instead with respect to overtime changes in such conditions. In the baseline models, overtime developments are captured by the differences in conditions obtaining across the seven waves of the ESS survey (between 2002 and 2014). But other specifications can do more to focus on overtime dynamics within a country. We consider two approaches. First, we added country dummies to the baseline specification, putting the emphasis on the “within-country” variation by controlling away country-specific conditions beyond the country-year substantive controls. Such specifications very much corroborate the results of our baseline models in Tables 2 and 3. A second, more aggressive, approach focuses on the possibility that voters focused on actual changes in conditions in a given year. To explore this possibility, we considered models including year-on-year differences in the aggregate (country-year) variables. Here we do not generally get significant results, except for the role of social welfare expenditure, which performs in line with the baseline models. We take this to mean that if one puts the emphasis fully on overtime dynamics, washing away the between-country variation as much as possible, then we see less support for our “dampening” hypotheses.

Eighth, an important alternative estimator that we considered is multilevel multinomial logit estimation, allowing a pooling of all the voting choices—such as for mainstream parties, for radical left, for radical right, and for nonvoting. Although such models are very computation-intensive and time-consuming, they allow judgment of whether we see links between economic well-being and radical right and radical left voting relative to the patterns of mainstream or nonvoting. We considered such models, first, with respect to the mainstream vote as a baseline and then considering other baseline options (e.g., radical left). The models corroborate our main results in Tables 2 and 3. For instance, relative to the mainstream vote as a baseline, subjective economic hardship (low economic well-being) spurs support for radical right parties but mainly in settings with lower unemployment rates and income inequality and higher social welfare spending and GDP. Relative to the same baseline (mainstream parties), the same macroeconomic conditions do not statistically significantly dampen the tendency of economic hardship to spur support for radical left parties. And economic hardship’s tendency to spur both radical right and left voting, relative to mainstream voting, tends in the multinomial logit specifications to be dampened by higher net immigration. Although this moderating effect of net migration is stronger for radical right than to have shown up in our baseline specification in Table 2, the multinomial approach does clearly corroborate the baseline support for hypotheses 2, 3, and 4.

Finally, we used the jackknife procedure and estimated our regression models for different subsets of the samples: We dropped the country-years one at a time to assess how sensitive the results are for outlying country-years. In general, the results indicate that the findings are robust. Only the interaction between well-being and the Gini coefficient in the case of radical right voting is sensitive to dropping some of the country-years. All other effects remain unaltered after dropping the country-years one at a time.

Conclusion

Radical left and right parties are increasingly successful in Europe, particularly among those who experience individual economic difficulties (Gomez et al., 2016; Kriesi et al., 2008, Kriesi et al., 2006; Lubbers et al., 2002; Ramiro, 2016; Rydgren, 2013; Werts et al., 2013). In this article, we have shown that the effect of individual economic well-being on radical voting depends significantly on national contexts. We formulated two contrasting (sets of) hypotheses. According to our deepening hypothesis, unfavorable national conditions deepen the relationship between individual hardship and radicalism: Those who are less well-off, vote for radical left and right parties particularly if macrolevel conditions are unfavorable, as national-level economic challenges make citizens even more uncertain about their individual economic position and yearn for radical challengers.

However, aggregate economic context could also have precisely the opposite moderating effect on the relationship between individual hardship and voting, and our exploration has distinguished two possible mechanisms underlying this dampening hypothesis. First, according to the “relative deprivation mechanism,” the less well-off living under favorable socioeconomic conditions might well benchmark their own economic hardship against the positive socioeconomic circumstances at the national level, feel that they do not get what they deserve, and feel more inspired to vote for a radical left or right party. The second “risk aversion mechanism” builds on the assumption that voting for a radical party is a risky strategy. Many radical parties lack office experience and propose radical changes with highly uncertain outcomes. Instead of making things better for those who experience economic hardship, these radical parties might through their inexperience and untested radicalism make things worse. Hence, less well-off citizens might only vote for a radical party under more favorable circumstances—if, in other words, these circumstances are relatively “safe” for radical experimentation. If these voters experience national external threats, they can become more risk averse and refrain from voting for a radical party.

We hypothesized, however, that radical right voting is risky under different circumstances than radical left voting. Radical right voting is risky for the less well-off if the socioeconomic circumstances in a country are unfavorable. After all, radical right parties are the issue owners of sociocultural issues like immigration; voters might well consider it doubtful if they are the ones to remedy unfavorable national economic conditions. Radical left voting is risky for the less well-off if national sociocultural conditions are unfavorable—that is, if the level of immigration is high (with high immigration levels the less well-off might fear that immigrants take their jobs). The reason is that radical left parties focus on socioeconomic issues and not sociocultural ones like immigration. And if they address the issue of immigration, they tend to hold rather multicultural positions.

Based on analysis of seven waves of ESS covering 21 countries, 27 radical right parties, and 22 radical left parties, we find no support for the deepening hypothesis and considerable support for the dampening hypotheses—yet with differences between voting for the radical right and left that are in line with our expectations based on the “risk aversion mechanism.” When it comes to radical right voting, someone’s well-being only exerts a negative effect if unemployment and inequality are low and GDP per capita and social welfare expenditure are high. The effect of well-being on radical right voting turned out not to be conditional on immigration, yielding a pattern consistent with the expected pattern of risk aversion for radical right voting.

When it comes to radical left voting, we found that the only moderator of the effect of well-being on voting behavior is immigration (and not unemployment, inequality, GDP per capita, and welfare expenditure): An individual’s socioeconomic position only leads to radical left voting if the level of immigration is low. Subsequent analyses indicated that this story only holds true regarding progressive radical left parties. That makes sense, because these parties tend to hold much more multicultural positions than more conservative radical left parties. Such patterns suggest clear party-color limits to the relative deprivation dampening hypothesis (H2): The tendency of economic hardship to spur radical left voting, as opposed to radical right voting, may have less to do with actual resentment or relative deprivation manifest in one’s suffering in a national context of prosperity. However, the patterns support a particular kind of risk aversion with respect to sociocultural immigration for radical left parties (H4).

To summarize, we found no support for the deepening hypothesis (H1), partial support for the relative deprivation dampening hypothesis (H2; only regarding radical right voting), and full support for the risk aversion dampening hypotheses (H3 and H4; both regarding radical right and radical left voting).

In general, our findings suggest that there is an important paradox of individual and aggregate economic well-being in the politics of radical voting: Although individual hardship spurs radical left and right voting, this is the case mainly when aggregate conditions are favorable, yielding a setting that is safe for radical experimentalism or positioning. At first sight, such corroboration of the dampening hypothesis seems to be at odds with the idea of economic voting where citizens vote in favor of the government if the economy is performing well but against the government if it is not. However, as we noted earlier, our study does not look at voting for government versus nongovernment parties but investigates voting for mainstream parties versus radical parties. This is an important difference that might well explain why the less well-off are more likely to vote radical under favorable rather than unfavorable aggregate-level conditions.

Our findings also indicate that this paradox works out differently for different kinds of radicalisms, for different kinds of aggregate circumstances, and also involving different mechanisms along the way. Although all our findings are particularly in line with the logic of risk aversion, this does not imply that we have shown that only the risk aversion mechanism takes place and that we can reject the logic of relative deprivation. After all, our findings vis-à-vis radical right voting are very much in line with the mechanism of relative deprivation—and this pattern comports with the view that particularly radical right populism is about resentments, where relative positioning is particularly important. In any event, we see the contrasting patterns for radical left and radical right voting with respect to the relative deprivation and risk aversion mechanisms as strong reasons for future research to focus on further unraveling these mechanisms.

Indeed, we believe that the study presented here has only begun to clarify when (less well-off) voters cast their votes for radical left and radical right parties. It could be expected that other macro variables, such as particular kinds of migration (Hainmueller & Hiscox, 2007), the media (Boomgaarden & Vliegenthart, 2007), parties themselves (Van Kessel, 2011), and political opportunity structures (Arzheimer & Carter, 2006) are of importance as well. Future studies might well want to incorporate such variables in their models. It could also be that individual economic hardship has implications moderated not only by national-level conditions but also or mainly by more meso- or local-level conditions, such as city or provincial socioeconomic fortunes. Although the data on such a level of analysis is harder to come by, looking at such conditions should clarify how an individual’s well-being or hardship plays out for political tastes as a function of various levels of his or her socioeconomic environment. Finally and perhaps most importantly, future study should investigate the causal mechanisms behind the “dampening effect” more closely—for example, by means of experiments in which the effects of informative cues about macrolevel economic conditions (unemployment, inequality, etc.) on party sympathy, and vote intention are investigated. Nonetheless, with the present article, we have brought research on radical voting an important step further: We now know a lot more about when and where unfavorable national contexts can dampen instead of deepen radical right and radical left among less well-off Europeans.

Footnotes

Declaration of Conflicting Interests

The authors declared no potential conflicts of interest with respect to the research,authorship,and/or publication of this article.

Funding

The authors received no financial support for the research,authorship,and/or publication of this article.

Supplemental Material

Supplementary material for this article is available online at the CPS website

Author Biographies

Matthijs Rooduijn is an assistant professor in the Department of Sociology at Utrecht University. His research focuses on the causes and consequences of the rise of populist and radical (left and right) parties.

Brian Burgoon is a professor in the Department of Political Science at the University of Amsterdam (UvA),and the director of the UvA’s Amsterdam Institute for Social Science Research (AISSR). His research focuses on the political conflicts surrounding economic globalization.

References

Arzheimer

(2009). Contextual factors and the extreme right vote in Western Europe, 1980–2002. American Journal of Political Science, 53, 259-275.

Arzheimer

Carter

(2006). Political opportunity structures and right-wing extremist party success. European Journal of Political Research, 45, 419-443.

Bale

Dunphy

(2011). In from the cold? Left parties and government involvement since 1989. Comparative European Politics, 9, 269-291.

Bergh

(2004). Protest voting in Austria, Denmark, and Norway. Scandinavian Political Studies, 27, 367-389.

Blekesaune

(2007). Economic conditions and public attitudes to welfare policies. European Sociological Review, 23, 393-403.

Blekesaune

Quadagno

(2003). Public attitudes toward welfare state policies: A comparative analysis of 24 nations. European Sociological Review, 19, 415-427.

Boomgaarden

H. G.

Vliegenthart

(2007). Explaining the rise of anti-immigrant parties: The role of news media content. Electoral Studies, 26, 404-417.

Bowyer

B. T.

Vail

M. I.

(2011). Economic insecurity, the social market economy, and support for the German left. West European Politics, 34, 683-705.

Catterberg

Moreno

(2005). The individual bases of political trust: Trends in new and established democracies. International Journal of Public Opinion Research, 18, 31-48.

10.

Crosby

(1976). A model of egoistical relative deprivation. Psychological Review, 83, 85-113.

11.

Dallinger

(2010). Public support for redistribution: What explains cross-national differences? Journal of European Social Policy, 20, 333-349.

12.

Dunphy

Bale

(2011). The radical left in coalition government: Towards a comparative measurement of success and failure. Party Politics, 17, 488-504.

13.

Eger

M. A.

(2010). Even in Sweden: The effect of immigration on support for welfare state spending. European Sociological Review, 26, 203-217.

14.

Ennser

(2012). The homogeneity of West European party families: The radical right in comparative perspective. Party Politics, 18, 151-171.

15.

Finseraas

(2009). Income inequality and demand for redistribution: A multilevel analysis of European public opinion. Scandinavian Political Studies, 32, 94-119.

16.

Golder

(2003). Explaining variation in the success of extreme right parties in Western Europe. Comparative Political Studies, 36, 432-466.

17.

Gomez

Morales

Ramiro

(2016). Varieties of radicalism: Examining the diversity of radical left parties and voters in Western Europe. West European Politics, 39, 351-379.

18.

Hainmueller

Hiscox

M. J.

(2007). Educated preferences: Explaining attitudes toward immigration in Europe. International Organization, 61, 399-442.

19.

Hakhverdian

Mayne

(2012). Institutional trust, education, and corruption: A micro-macro interactive approach. The Journal of Politics, 74, 739-750.

20.

Han

K. J.

(2016). Income inequality and voting for radical right-wing parties. Electoral Studies, 42, 54-64.

21.

Hawkins

K. A.

(2010). Venezuela’s Chavismo and populism in comparative perspective. Cambridge, UK: Cambridge University Press.

22.

Hooghe

Marks

Wilson

C. J.

(2002). Does left/right structure party positions on European integration? Comparative Political Studies, 35, 965-989.

23.

Ivarsflaten

(2005). The vulnerable populist right parties: No economic realignment fuelling their electoral success. European Journal of Political Research, 44, 465-492.

24.

Ivarsflaten

(2008). What unites right-wing populists in Western Europe? Comparative Political Studies, 41, 3-23.

25.

Jackman

R. W.

Volpert

(1996). Conditions favouring parties of the extreme right in Western Europe. British Journal of Political Science, 26, 501-521.

26.

Kitschelt

(1995). The radical right in Western Europe. Ann Arbor: University of Michigan Press.

27.

Kriesi

Grande

Lachat

Dolezal

Bornschier

(2008). West European politics in the age of globalization. Cambridge, UK: Cambridge University Press.

28.

Kriesi

Grande

Lachat

Dolezal

Bornschier

Frey

(2006). Globalization and the transformation of the national political space: Six European countries compared. European Journal of Political Research, 45, 921-956.

29.

Lewis-Beck

M. S.

Stegmaier

(2000). Economic determinants of electoral outcomes. Annual Review of Political Science, 92, 759-774.

30.

Lubbers

Gijsberts

Scheepers

(2002). Extreme right-wing voting in Western Europe. European Journal of Political Research, 41, 345-378.

31.

Lubbers

Scheepers

(2007). Euro-scepticism and extreme voting patterns in Europe: Social cleavages and socio-political attitudes determining voting for the far left, the far right and non-voting. In Loosveldt

Swyngedouw

Cambré

(Eds.), Measuring meaningful data in social research (pp. 71-92). Leuven, Belgium: Acco.

32.

Lucassen

Lubbers

(2012). Who fears what? Explaining far-right-wing preference in Europe by distinguishing perceived cultural and economic ethnic threats. Comparative Political Studies, 45, 547-574.

33.

March

(2011). Radical left parties in Europe. New York, NY: Routledge.

34.

March

Rommerskirchen

(2015). Out of left field? Explaining the variable electoral success of European radical left parties. Party Politics, 21, 40-53.

35.

Mau

Burkhardt

(2009). Migration and welfare state solidarity in Western Europe. Journal of European Social Policy, 19, 213-229.

36.

Mudde

(2007). Populist radical right parties in Europe. Cambridge, UK: Cambridge University Press.

37.

Ramiro

(2016). Support for radical left parties in Western Europe: Social background, ideology and political orientations. European Political Science Review, 8, 1-23.

38.

Rooduijn

Akkerman

(2017). Flank attacks: Populism and left-right radicalism in Western Europe. Party Politics, 23, 193-204.

39.

Rooduijn

Burgoon

B. M.

van Elsas

E. J.

van de Werfhorst

H. G.

(forthcoming). Radical Distinction: Support for Radical Left and Radical Right Parties in Europe. European Union Politics.

40.

Runciman

W. G.

(1966). Relative deprivation and social justice. London, England: Routledge.

41.

Rydgren

(Ed.). (2013). Class politics and the radical right. New York, NY: Routledge.

42.

Rydgren

Ruth

(2013). Contextual explanations of radical right-wing support in Sweden: Socioeconomic marginalization, group threat, and the halo effect. Ethnic and Racial Studies, 36, 711-728.

43.

Schneider

S. L.

(2008). Anti-immigrant attitudes in Europe: Outgroup size and perceived ethnic threat. European Sociological Review, 24, 53-67.

44.

Schumacher

Rooduijn

(2013). Sympathy for the “devil”? Voting for populists in the 2006 and 2010 Dutch general elections. Electoral Studies, 32, 124-133.

45.

Semyonov

Raijman

Gorodzeisky

(2006). The rise of anti-foreigner sentiment in European societies, 1988-2000. American Sociological Review, 71, 426-449.

46.

Smith

H. J.

Pettigrew

T. F.

Pippin

G. M.

Bialosiewicz

(2012). Relative deprivation: A theoretical and meta-analytic review. Personality and Social Psychology Review, 16, 203-232.

47.

Solt

(2014, March). Standardized World Income Inequality Database v4.1. Retrieved from fsolt.org

48.

Sperber

(2010). Three million Trotskyists? Explaining extreme left voting in France in the 2002 presidential election. European Journal of Political Research, 49, 359-392.

49.

Stegmueller

(2013). How many countries do you need for multilevel modeling? A comparison of Bayesian and frequentist approaches. American Journal of Political Science, 57, 748-761.

50.

Van der Brug

Fennema

Tillie

. (2000). Anti-immigrant parties in Europe: Ideological or protest vote? European Journal of Political Research, 37, 77-102.

51.

Van der Brug

Fennema

Tillie

. (2005). Why some anti-immigrant parties fail and others succeed. Comparative Political Studies, 38, 537-573.

52.

Van der Meer

. (2010). In what we trust? A multi-level study into trust in parliament as an evaluation of state characteristics. International Review of Administrative Sciences, 76, 517-536.

53.

Van der Meer

te Grotenhuis

Pelzer

. (2010). Inﬂuential cases in multilevel modeling: A methodological comment. American Sociological Review, 75, 173-178.

54.

Van Kessel

. (2011). Explaining the electoral performance of populist parties: The Netherlands as a case study. Perspectives on European Politics and Society, 12, 68-88.

55.

Visser

Lubbers

Kraaykamp

Jaspers

(2014). Support for radical left ideologies in Europe. European Journal of Political Research, 53, 541-558.

56.

Walgrave

De Swert

(2004). The making of the (issues of the) Vlaams Blok. Political Communication, 21, 479-500.

57.

Walker

Pettigrew

T. F.

(1984). Relative deprivation theory: An overview and conceptual critique. British Journal of Social Psychology, 23, 301-310.

58.

Weakliem

D. L.

(2002). The effects of education on political opinions: An international study. International Journal of Public Opinion Research, 14, 141-157.

59.

Werts

Scheepers

Lubbers

(2013). Euro-scepticism and radical right-wing voting in Europe, 2002–2008: Social cleavages, socio-political attitudes and contextual characteristics determining voting for the radical right. European Union Politics, 14, 183-205.

60.

World Development Indicators. (2016). World Bank Databank Net migration statistics. Retrieved from http://data.worldbank.org/indicator/SM.POP.NETM

61.

Zhirkov

(2014). Nativist but not alienated: A comparative perspective on the radical right vote in Western Europe. Party Politics, 20, 286-296.

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The Paradox of Well-being: Do Unfavorable Socioeconomic and Sociocultural Contexts Deepen or Dampen Radical Left and Right Voting Among the Less Well-Off?

Abstract

Keywords

Radical Parties and Radical Voting

Unfavorable Contexts: Deepening or Dampening Radical Voting Among the Less Well-Off?

Data and Measurement

Findings

Robustness Checks and Extensions

Conclusion

Footnotes

Declaration of Conflicting Interests

Funding

Supplemental Material

Author Biographies

References

Supplementary Material