Economic voting in Spain: A 2000 panel test

Electoral Studies 29 (2010) 210–220

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Economic voting in Spain: A 2000 panel test Marta Fraile a, *, Michael S. Lewis-Beck b a b

Research Fellow at Consejo Superior de Investigaciones Cientı´ficas (CSIC, IPP), Madrid, Spain Department of Political Science, University of Iowa, United States

a r t i c l e i n f o

a b s t r a c t

Article history: Received 13 April 2009 Received in revised form 21 December 2009 Accepted 22 January 2010

Considerable research shows the economy matters for voters. But that view has come under attack, with revisionists arguing that it matters little. This dissenting view fits the Spanish case well, where reigning research finds virtually no economic voting exists. We argue against the revisionist view, suggesting that conclusion stems largely from methodological limitations in its supporting cross-sectional survey analyses. Given the causality question these analyses raise, particularly in the context of likely endogeneity, a panel analysis is called for. We examine the most recent available panel survey, from the 2000 general election, estimating fully specified multinomial logit models. We find strong economic effects. Spain appears, after all, to have an electorate capable of holding the government economically accountable, at least in this instance. Ó 2010 Elsevier Ltd. All rights reserved.

Keywords: Economic voting Endogeneity Ideological voting Spanish elections

1. Introduction In the field of political behavior, economic voting has been heavily studied. By one recent estimate, there are over 400 books and articles on the subject (Lewis-Beck and Stegmaier, 2007). The overwhelming conclusion is that economic evaluations generally have a statistically significant, and often substantively strong, impact on government support. [See reviews in the following as well: Anderson (1995, chp.3); Nannestad and Paldam (1994); Norpoth (1996); Duch and Stevenson (2008).] But, against this conventional wisdom presses a rising tide. The revisionists, as they have been called, argue that citizens’ economic assessments are little more than rationalizations of their prior partisanship (Anderson, 2007; Anderson et al., 2004; Evans and Andersen, 2006; Ladner and Wlezien, 2007; Wlezien et al., 1997). Individuals with an attachment to the incumbent will perceive the national economy more positively. These electors, once decided to vote for the

* Corresponding author. Tel.: þ34 91 6022946; fax: þ34 91 6022971. E-mail addresses: [email protected] (M. Fraile), [email protected] (M.S. Lewis-Beck). 0261-3794/$ – see front matter Ó 2010 Elsevier Ltd. All rights reserved. doi:10.1016/j.electstud.2010.01.003

incumbent on partisan grounds, look for ex-post ways of rationalizing their electoral decisions, in the claim that the economy is doing well. In other words, there is no true, independent effect of economic thinking on the vote. The revisionist argument has special relevance for the Spanish literature where, in contrast to that for other leading Western European democracies, a version of it has long held sway. Indeed, for the Spanish case, the winning argument has been that partisan ideology more or less completely shapes economic evaluations. Voters see only their ideology, not the real economy. Thus, governments are not held accountable for the actual economic state of things. Empirically, this makes Spain an exceptional case, in the European context. Normatively, this has undesirable implications, suggesting a democratic deficit. The government, free of sanctions from the voters, simply follows the economic course of its own choosing. Recent research has shown that the revisionist argument, as persuasive as it might seem theoretically, receives challenge under econometric testing for causality in a panel setting (Lewis-Beck, 2006; Lewis-Beck et al., 2008). Despite a demanding analysis, the independent effect of the economic vote emerges strongly. But one immediate difficulty with this conclusion is its empirical confinement to

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simpler political systems – US, UK, Canada – Anglo-Saxon democracies with only two (or perhaps two-and-onehalf) major parties. The more complex Spanish presidential-parliamentary system, with its multiple major parties, might not reveal significant economic voting, even given ideal test conditions. Here we take on this apparent Spanish exceptionalism. Below, we first review the Spanish case for ‘‘no economic vote.’’ As solid as that case may appear, it is vulnerable on at least two counts. First, it rests on crosssectional survey analysis. Second, the potential endogeneity problem with respect to ideology and economic evaluation has not been previously addressed. Considering the first issue, cross-sectional analysis offers a weak foundation for causal inference. We know that when X causes Y, change takes place over time. However, in a cross-sectional survey, X and Y are measured at virtually the same moment, so temporality and causal sequence can only be inferred. This difficulty made Lewis-Beck and Stegmaier (2000, 195) call for ‘‘panel data to explore the temporal dynamic of individual economic voting.’’ Considering the second issue, the exogeneity of ideological partisanship with respect to economic evaluation should not simply be assumed. Ideology may well influence economics, but economics may also influence ideology. If this reciprocal causation actually exists for Spain, the variables are really endogenous, and ordinary estimation of effects becomes biased. To deal with these critical issues, we go on to utilize panel survey data, appropriately adjusting it for any endogeneity problem. For Spain, there are two sets of panel data available, for the 1993 and the 2000 general elections. The former does not work, because it fails to contain adequate batteries of relevant items, measured more than once. Fortunately, the latter is up to the task, as well as being more current. The 2000 General Election Panel Survey, carried out by the Center for Sociological Research (CIS) in two waves, is based on a representative probability sample of the adult Spanish population. (More details are offered below). 2. The Spanish political system and the 2000 elections As a parliamentary democracy, Spain typically has several parties competing in any given election. In a multiparty context, electors might encounter more difficulty in following the economic logic of voting, as compared to a presidential system. Multiparty systems often involve coalition, or at least non-majoritarian, governments. Given these circumstances, it can be more difficult for voters to hold the government responsible for economic outcomes. Since the beginning of the process of democratic consolidation, there have only been two main parties with a strong chance of holding the reins of central government – the Socialist party (PSOE) and the Popular Party (PP). The Socialists were in power across four consecutive elections, 1982–1996, with the right-wing Popular Party serving as the main opposition. At the beginning of its opposition role, its credibility as a democratic party ready to govern Spain was low. However, PP electoral support improved, while the Socialist percentage vote share steadily decreased over time: 48.11% in 1982 (202 seats), 44.06% in 1986 (184 seats),

211

39.60% in 1989 (175 seats), 38.78% in 1993 (159 seats). [Department of the Interior: http://www.elecciones.mir.es/ MIR/jsp/resultados/index.htm]. Across the democratic period, the continual return of the Socialists (the PSOE) to power might suggest, at first blush, that there was no significant economic voting (Hamann, 2000). These aggregate figures, however, do not necessarily mean an absence of economic voting at the individual level. Indeed, the observed decline in the PSOE winning margin over time is compatible with the growing presence of an economic voting effect. (We return to this point later.) Moreover, the PP finally managed to defeat the ruling PSOE in 1996, and entered the 2000 contest with a record of very good economic performance (of which we say more below). Besides the vibrant economy, the PP was advantaged by the internal divisions of the PSOE, brought about by the unexpected resignation of their General Secretary, Felipe Gonzalez. A charismatic leader of the Socialists since 1974, he designated Alumina as his heir. Nevertheless, Alumina lost the primary elections for that post, held within the party. Furthermore, Borrell, the actual winner of these primaries, could not hold the support of the party majority, and finally resigned. Alumina was left, after all, as the final candidate for the 2000 General Election, in which the PSOE obtained it worst results since 1977. 3. Spain within the economic voting literature There have been numerous investigations of economic voting in the democracies of Western Europe, both singleand multi-nation. (The most current literature review here is Lewis-Beck and Stegmaier, 2007). In these examinations, the classic reward-punishment paradigm of retrospective economic voting has been generally sustained (Fiorina, 1981; Lewis-Beck, 1988). When electors are satisfied with the economy, they give support to the incumbent, but when they are dissatisfied they withdraw support. This economic voting hypothesis has been tested with Spanish elections (Fraile, 2005; Gonza´lez, 2002; Lancaster and Lewis-Beck, 1986; Maravall and Przeworski, 2001). Taken together, the results they report are inconclusive. In one of the most comprehensive studies, spanning the period 1980–1995, Maravall and Przeworski (2001) find that electors’ ideology tends to neutralize the effect of their economic judgments on the incumbent vote. Even when voters appear to respond to economic conditions, such responses are ideologically driven. Overall, they decide that Spanish electors settle on an interpretation of economic conditions that allows them to vote according to their ongoing ideological preferences. Specifically, in a multinomial logit model they estimate vote intention as a function of socio-demographic variables, ideology, and subjective national economic evaluations. (They pool 63 crosssectional surveys, replicating the same analysis for five different phases of the sixteen-year period). Economic voting, they continue, is severely limited by political loyalties and ideology (Maravall and Przeworski, 2001, 54). The causal arrow does not go from electors’ views about economic performance to their vote; rather, exactly the reverse. They explain these results by arguing

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that the social and ideological roots of party preferences in Spain are consistent, with the majority of the political parties competing in general elections having a long history. ˜ ol) [For instance, the PSOE (Partido Socialista Obrero Espan was created in 1879; the PNV (Partido Nacionalista Vasco) in ˜ a) in 1921, the 1895; the PCE (Partido Comunista de Espan ERC (Esquerra Republicana de Catalunya) in 1931.] The electors, then, rely on this historical memory, further accounting for the ideological contamination of economic voting in Spain. In the words of Maravall and Przeworski (2001, 71) themselves, ‘‘Causality seems to have been often reversed: partisan allegiances rooted in social and political conditions influenced the interpretations of the economy. Once class and ideology mediated, people interpreted the same economic conditions in different ways and could thus maintain their ideological and partisan loyalties in the face of changing economic circumstances.’’ We counter that this argument itself has limitations, since a number of important parties were created during the democratic transition, and so lack a long history. That is the ´tico (UCD), the center-right case with Union de Centro Democra party founded in 1977, and in power from then to 1982. Also, Alianza Popular (AP, later on Partido Popular, PP) founded in ´ (CIU), founded in 1978. These 1976, or Conve`rgencia i Unio founding dates suggest that economic voting might be less constrained than supposed by historic loyalties. Fraile (2005) partakes of this view, in her own analysis of a comparable, still longer period, 1979–1996. She employs cross-sectional surveys carried out immediately prior to each general election (since there were no postelection surveys that contained questions about the state of the economy). In a logit analysis, she tests the effects of subjective national economic evaluations, in comparison to the effects of ideology or class. After the transition to democracy (beginning with the 1979 general elections), the ideological component of the vote, important as it was, tended to diminish, while the economic component tended to increase. She decides that ‘‘in light of the results obtained here, it can safely be concluded that the Spanish electorate does react to changing policy outcomes, social conditions and economic circumstances. Moreover, citizens react to governments’ economic policies and to the perceived results of these policies.’’ (Fraile, 2005, 222). In the contrasting results of these comprehensive studies, then, we see most clearly two opposing points of view. Neither of them, of course, has been fully tested against the 2000 general elections, using individual election survey data. This seems unfortunate, given the supposed salience of the economy at the time, at least by certain accounts. Under the incumbent conservative government of the Partido Popular (PP) the economy performed well. The GDP real growth rate increased from 2.4 percent (1996) to 5.0 percent (2000). The total jobs growth rate rose from 1.7 percent (1996) to 5.1 percent (2000). Finally, the unemployment rate fell dramatically, from 17.8 (1996), to 11.1 (2000). [UNECE, United Nations Economic Commission for Europe; www.unece.org/stats/dta.htm ] Moreover, the economy was a remarkably prominent issue during the electoral campaign, embodied in the famous slogan of prime minister Aznar’s campaign staff, ‘Spain is doing well.’

Nevertheless, despite these impressive economic statistics, published scholarly investigations of the 2000 election underplay economic voting, relying instead on traditional explanations of the ideological component of the vote, coupled with social structural determinants such as religion (see, respectively, Torcal and Medina, 2002; and Calvo and Montero, 2002). The last set of authors sums up this view well: ‘‘The recent division among religiosity and ideology has forced the former to impinge on the vote in a similar way as ideology used to do in the past.’’ (Calvo and Montero, 2002). A dissenting voice to this conclusion comes from Gonza´lez (2002), who explains the 2000 results as a consequence, after all, of economic voting. However, his findings exact no privilege over the others because they too are based on crosssectional evidence. We take issue with the dominant interpretation – the lack of an economic vote in Spain - especially as applied to the 2000 context. The 2000 General Election Panel Survey, conducted by CIS and composing a national probability sample, took place in two waves. In the first wave, face-toface interviews were carried out in February one month before the general election, with N ¼ 24,040 adults, randomly sampled (within strata of the Spanish regions). The second wave was a random sample from the first wave sample, constituting a representative national sample. The interviews took place after the general election, from March 10 to the end of April, and were again face-to-face, for a total N ¼ 5283 (interviewed in both waves). Here we utilize this sample interviewed in both waves, dubbing the first the preelection wave, the second the post-election wave. For full details about the sampling procedures, the technical report of the CIS can be consulted (http://www.cis.es). 4. Economic evaluation and political Ideology: initial panel evidence As Lazarsfeld (1955) reminded us long ago, a prime requisite for causal inference is that the independent variable occurs before the dependent variable. More immediately, Lewis-Beck and Stegmaier (2000, 195) comment, ‘‘When individuals are surveyed at two or more points in time.the variation is more likely to be causal.’’ We exploit the strength of this notion, examining declared vote at time t as a function of economic evaluation and political ideology at t  1. The time lag of the pre- and post-election 2000 survey is not long, at two months, but appears sufficient for our purposes. In the case at hand, the lag is great enough that the respondents’ answers in the two waves do exhibit noticeable differences. For example, political ideology and economic evaluation measured at time t  1 are correlated with the time t measure, respectively, at r ¼ .67 and r ¼ .44. While these correlations show some stability, they also evidence change. In sum, the two-month time lag between the surveys serves our research ends, since there is relevant change of both ideology and economic evaluation across the period. Our analysis strategy is to estimate voting models with a standard specification, one that closely follows that of Maravall and Przeworski (2001), who examine the

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Table 1 Preliminary panel estimates (Multinomial logit).a Vote for PP versus PSOE (1) Ideology (t) b Ideology (t-1) Economy (t) Economy (t-1) Age Male Education Social class: Service class White collar Skilled manual worker Unskilled manual worker

(2)

2.2

(.08)

**

.92

(.09)

**

-.006 .13 .05

(3)

1.51

(.06)

**

.64 -.003 .15 .007

(.07) (.003) (.11) (.05)

**

(.003) (.12) (.05)

.20 .07 -.77

(.23) (.21) (.19)

-.02 .005 -.56

(.21) (.18) (.17)

Urban Intercept

-.44 -.13 -8.89

(.19) (.12) (.49)

-.57 -.07 -5.7

(.38) (.11) (.38)

Pseudo-R-sq. N

.25 3432

** * **

.18 3376

** *

1.48 1.11

(.06) (.08)

-.006 .06 -.01

(.003) (.11) (.04)

.05 .13 -.86

(.21) (.19) (.17)

-.48 -.06 -7.38

(.17) (.11) (.44)

** **

** ** **

.19 3389

** ¼ Statistical significance, .01; * ¼ Statistical significance, .05; two-tailed tests. Figures in parentheses are standard errors. a Dependent variable ¼ 1 Vote for the incumbent (PP), 2 Vote for PSOE, 3 Vote for IU, 4 Else, 5 Abstention; Only the coefficients among the two main parties are shown. The complete multinomial results (for all the categories of the dependent variable) are given in Appendix 2. (See the corresponding estimation for each table and equation.) b Independent variables: Ideology ¼ 2 if extreme left to 0 for center to +2 if extreme right; Economy ¼ 2 for very bad national economic conditions to 0 for regular national economic conditions to +2 for very good national economic conditions; Age ¼ in years; Male ¼ 1 for male, 0 for female; Education ¼ from lowest (1) to highest (6); For Social Class self employed is the category of reference; urban ¼ 1 for areas more than 50 thousands inhabitants, 0 for areas less than 50 thousand inhabitants. Source: CIS2382 and 2384 (pre- and post-2000 elections survey).

determinants of socio-demography, ideology, and national economic perceptions. This parallel conceptual specification permits cautious comparison of findings across the two modeling efforts. Further, since the specification is relatively parsimonious, it keeps the main findings in view, as the testing requirements increase. Also, it is a conservative strategy, working against the economic voter hypothesis. That is to say, if variables such as party identification and religious practice are excluded, the ideology variable has more opportunity to claim bigger effects. (Practically speaking, it is impossible to include party identification in any case, since it is not properly measured in these surveys. As well, while religiosity is measured, once included in the specification it fails to alter the substantive conclusions.) The essential equation that we will estimate, with multinomial logit (MNL), reads thusly, Vote ¼ f ðsocio  demographics; political ideology; economic evaluationÞ: (1) Vote (V) is a categorical variable, scored as follows: 1 ¼ PP (the incumbent party, conservative); 2 ¼ PSOE (the main opposition party, social democrat); 3 ¼ IU (the United Left, at the left of the PSOE); 4 ¼ other (all remaining oppositions, including nationalist parties); 5 ¼ abstention. The socio-demographic (S) variables are as follows – age, gender, education, class, rural/urban residence – coded as noted at the foot of Table 1. The economic evaluation (E) measure, a standard fivepoint sociotropic retrospective judgment on the nation’s general economic situation, goes from ‘‘very bad’’ (2) to

‘‘very good’’ (þ2). Other standard economic perception measures - sociotropic prospective, egotropic retrospective, egotropic prospective – are not available, since they were not asked in both waves (Lewis-Beck, 1988). Lastly, political ideology (I) composes a five-point self-placement scale, from ‘‘left’’ to ‘‘right’’ where 2 is the most extreme left score, and þ2 the most extreme right score, with ‘‘0’’ in the middle. The variable of ideology has a long and important place in Spanish election surveys. Indeed, this is one of the few variables that has always been included in post-electoral surveys in Spain (see http://www.cis.es). Virtually without fail, it has been a single-item ideological self-placement scale, with respondents locating themselves from left to right. For example, in 2000, a ‘‘1’’ on the continuum represents the extreme left score, where as ‘‘10’’ represents the extreme right score. As a practical matter, it makes little difference whether the ten-point or five-point scale is used, in that either renders essentially the same correlations with the vote. A value of using the five-point scale, then, lies in its comparability to the five-point metric of the economic evaluation. Once our two key variables – E and I - are both coded to the same metric, we can compare their coefficients directly, in order to assess the relative magnitude of their effects. (In the footnote, we further discuss a simulation which directly calculates the change in the vote probability, as a function of changes in the E or I scores. This simulation approach leads to the same conclusions about the relative importance of effects from political ideology and economic evaluation, as does the direct comparison of coefficients approach,

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which we emphasize here simply for purposes of exposition). 1 The static equation below measures all variables conventionally, at time t. Vt ¼ f ðIt; Et; StÞ

(2)

A dynamic equation, in contrast, measures the variables at different times. Vt ¼ f ðIt  1; Et  1; StÞ:

(3)

The conventional model, of equation (2), is estimated with MNL and appears in column 1, Table 1. As can be seen, the reference category is PP versus PSOE, controlling for the other vote categories. This stands as the choice of paramount interest, since PP is the incumbent, and the PSOE is the main opposition. (The tables for the other vote categories, with PP remaining the reference, are not included because of space concerns; however, they are available upon request. The substantive conclusions about economic voting here, and later in the article, do not alter when these other categories are examined). Observe that the logit coefficients of both economics and ideology are statistically significant, and in the expected direction. That is, voters more ideologically on the right are more likely to support the incumbent conservative PP, as are voters who think the economy is doing better. Comparing the magnitude of the coefficients, the effect of political ideology appears something over twice as strong as that of economic evaluation; i.e., 2.2/.92 ¼ 2.39. But, in the revisionist view, this comparison is highly misleading, for it falsely magnifies the impact of economic evaluations, which in the Spanish case are immediate projections of the voter’s own political ideology. As a possible remedy, suppose we take advantage of the dynamic nature of the data, and lag the variables back in time, allowing them to crystallize prior to the vote decision. This exercise is carried out in column 2, Table 1, where both economics and ideology are measured at time t  1. One sees the effect of economics does decline. But so does the effect of ideology, and to about the same extent, leaving the coefficients ratio virtually unchanged (1.51/.64 ¼ 2.36).

1 In the original survey, the economics (E) item has five categories (coded here 2, 1, 0, þ1, þ2), while ideology (I) has ten categories (1– 10). To make the metrics for the two variables comparable, we coded ideology thusly: 2 ¼ (1, 2), 1 ¼ (3, 4), 0 ¼ (5), þ1 ¼ (6, 7), þ2 ¼ (8, 9, 10). Besides comparing the raw MNL coefficients of E and I directly, we might examine the change in probability of voting incumbent (versus opposition), as the values of these two independent variables change. Hence, we simulated the magnitude of the effect of changing values of ideology and economy (from their lowest value: 2 to their highest value: þ2), holding constant the rest of the equation variables at their sample mean. The substantive conclusions about economic voting extracted from Tables 1–3, based on a comparison of the raw coefficients, remain the same. Take as an example equation (3), Table 3 (perhaps the most demanding of the estimations). The probability of a PP vote changes from .16 to .79, as Ideology changes from 2 to þ 2. And, as the Economic Evaluation changes from 2 to þ2, the probability of a PP vote changes from .10 to .73. In other words, comparable changes in Ideology and Economic Evaluation produce comparable changes in the probablity of an incumbent vote. This simulation exercise, and those from the other equations in these tables, are summarized in Appendix, Table 1.

Why does this occur? The difficulty could stem from the fact that both ideology and economics are measured at the same time, in the pre-election survey. Ideological thinking may actually shape economic thinking, e.g., someone on the right might be more inclined to think that the economy under the PP was better than it was. Indeed, such endogeneity is precisely the mechanism distorting the voter’s economic perception, according to the revisionist hypothesis. To remove this ideological coloration from the voter’s immediate economic evaluation, ideology might logically better be measured at some distance prior. This possibility is tested in column 3, Table 1, with ideology measured in the pre-election wave, and economics measured in the post-election wave. If the revisionist hypothesis is correct, the economics coefficient should drop considerably, perhaps even to zero, since the contamination from ideology is lessened. The economics coefficient does not, however. Instead, it actually rises, to 1.11, while the ideology coefficient drops a bit, to 1.48. The resulting ratio (1.48/1.11 ¼ 1.33) marks a decided improvement in the relative importance of economics. 2 Overall, these preliminary results, by exploring the simple time dynamics of the relationship among these variables, point away from the notion that political ideology contaminates economic evaluations. On the basis of this initial testing, economic effects appear relatively large and persistent. 5. Economics and Ideology: solving the bias problem The panel analysis thus far has placed our causal inferences on firmer ground, by flexibly incorporating the temporality dimension. However, there are other inference difficulties to be faced. First, the models estimated here (along with the traditional cross-sectional models), implicitly hold that the ideology variable is exogenous, i.e., stable, enduring, unchanging. Clearly, this is the dominant characterization in the Spanish case, as the cited literature reveals, with ideology assumed to operate unidirectionally on economics. But suppose ideology is actually endogenous, i.e., it is unstable, malleable, changing. Certainly, this is the implication of the great variation in electors’ ideology over time, as reported by Torcal and Medina (2002). In such a situation, its effect, as estimated in Table 1, is biased upward. A source of this endogeneity might be the shortterm economic evaluations themselves, operating to alter a voter’s ideological position. For instance, citizens with

2 Because the panel data come in two waves, it becomes possible to specify different lag structures. Measuring both I and E at time t reproduces the cross-sectional design (see Column 1, Table 1). Lagging both, back to t  1, takes advantage of dynamics, and puts both variables theoretically on the same time footing causally (see Column 2, Table 2a,b). However, the revisionists argue that measuring both at the same time point actually biases estimation in favor of economic effects, because contemporaneous economic evaluations are actually transmitting much from ideology (i.e., the endogeneity problem). In that case, the bias would be reduced by lagging ideology but not economy (i.e., economics now is less under the influence of ideology because it is not measured contemporaneously). Still, some might prefer to avoid differential time lags. For that reason, we rely on the t  1 time lag for both, in Table 2a and in our final, most rigorous tests, in Table 3.

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a centrist ideology might move to a more conservative ideology, when they observe economic prosperity under a PP incumbency. In sum, there exists the possibility of ideology influencing economics but, in turn, economics influencing ideology. This poses the classic problem of simultaneous equation bias, and ordinary estimation procedures fail. Second, with panel data, there is the inevitable difficulty of correlated errors, which also biases estimates, unless corrective steps are taken (Finkel, 1995, chp.5). The problem stems from the time series nature of the data, where influences at an earlier time continue to operate at a later time. Let an ensemble of variables acting on ideology at t  1 be expressed as an error term, et  1, and an ensemble of variables acting on the vote at time t expressed as another error term, et. Suppose one of these variables, political participation level, is represented in both et  1 and et, because it influences both ideology and vote. Then, the error terms will necessarily be correlated, r et, et  1 > 0. It is not hard to imagine other such variables nested in these error terms. A remedy is to make the endogenous variables, such as ideology, effectively exogenous. Thus, in the estimates of Table 1, there are two likely sources of bias – from endogeneity and from correlated errors. A standard econometric correction is instrumental variables estimation, in two stages. [For a clear treatment of the method, see Wooldridge (2006)]. During the first stage, endogenous variable Y, say ideology, is regressed against true exogenous variables, rendering an instrumental variable, Y0. During the second stage, this Y0 replaces original Y in the equation to be estimated. The estimate for the Y0 coefficient will now have the desirable statistical property of consistency, because the variable Y has become exogenous (by construction). In political science, perhaps the most well-known example of instrumental variables estimation, as applied to panel data, is in Fiorina (1981, p. 170). He builds an instrument for party identification at time t  1, in order to estimate its effect on the vote at time t. In selecting exogenous variables for this construction, Fiorina (1981, 229– 230) focuses on two types: socio-economic variables and basic attitudes. We begin with a similar strategy, in our search for exogenous variables to build the ideology instrument. Specifically, we select the following variables, tapping, respectively, the partisan, evaluative, social cleavage, and socialization components of citizens’ ideological dispositions: PP support in 1996, consumption of political information, education, social class, church attendance, gender, age, and parents’ ideology. Thus, in the first stage, our instrumental variable for ideology, I0 , is constructed by regressing (OLS) the original ideology variable, I, on these variables. The details of the coding, and the results of the estimation of these ideology instruments (under different conditions of exogeneity) are available at http://www.ipp. csic.es/Pi-esp/fraile.html. In the second stage, this instrumental variable, I0 , is substituted for I in the original equation, which is then reestimated with MNL. In column 1, Table 2a, are these reestimates. The specification of this equation mirrors that

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Table 2a Exogenous Ideology (lagged), Economic Perception (lagged), and the Vote: Panel Tests (Two-Stage, Multinomial Logit Estimates).a (1) Vote for PP versus PSOE a Economy (t-1) b Ideology’(t-1, broad x)c Ideology’ (t-1, mod x) Ideology’ (t-1, strict x) Controls (not shown) d Intercepts (not shown) Pseudo-R-sq. N

(2)

(3)

.55 (.06) ** .87 (.06) ** .87 3.02 (.12) ** 2.31 (.12) ** 1.7

.21 3991

.12 3991

(.05) **

(.14) **

.09 4002

Source: CIS2382 and 2384 (pre and post 2000 elections survey).

of column 2, Table 1, taking full advantage of the temporal dimension of the panel, and including the same control variables (whose coefficients are not reported here, for clarity of presentation). The coefficient of the economics variable drops a bit (from .64 to .55), while that of ideology doubles (from 1.51 to 3.02). What is to be made of this result? At first blush, the strength of ideology appears greater than previously thought. However, as with any complex estimation from non-experimental data, before drawing conclusions the underlying assumptions of the analysis must be verified.

Table 2b Exogenous Ideology (lagged), Economic Perception (not lagged), and the Vote: Panel Tests (Two-Stage, Multinomial Logit Estimates).a (1) Vote for PP versus PSOE a Economy (t) b Ideology’(t-1, broad x)c Ideology’ (t-1, mod x) Ideology’ (t-1, strict x) Controls (not shown) d Intercepts (not shown) Pseudo-R-sq. N

(2)

(3)

1.07 (.08) ** 1.25 (.07) ** 1.28 (.07) ** 2.99 (.12) ** 2.23 (.12) ** 1.72 (.14) **

.22 4017

.13 4017

.11 4028

** ¼ Statistical significance, .01; * ¼ Statistical significance, .05; one-tailed tests. Figures in parentheses are standard errors. a Dependent variable ¼ 1 Vote for the incumbent (PP), 2 Vote for PSOE, 3 Vote for IU, 4 Else, 5 Abstention; Only the coefficients among the two main parties are shown. The complete multinomial results (for all the categories of the dependent variable) are available on request. b The economy variable is measured as in Table 1. c The instrumental variables: Ideology’ t  1 (strict x) ¼ a ‘‘strictly’’ exogenous ideology variable, an instrument constructed from the following variables: Objective social class (2. Service class, 3 White collar, 4 Skilled manual worker, 5 Unskilled manual worker) where self employed is the category of reference; Religion (from 0 for not attending church at all to 5 times a week); Age ¼ in years; Gender (1 for male and 0 for female); Education ¼ from lowest (1) to highest (6); Ideology’ t  1 (mod x) ¼ a moderately exogenous ideology variable, an instrument constructed from the foregoing variables in the strict construction, plus parents’ ideology (as a proxy for political socialization): 1. left, 2. center, 3. right, 4 no information (center is the category of reference); and consumption of political information in newspaper (from 1: every single day to 6 for never); Ideology’ t  1 (broad x) ¼ a broadly exogenous party identification variable, an instrument constructed from all the foregoing variables, plus reported vote to PP (the incumbent in 2000 elections) in 1996 (1 ¼ voted for the incumbent PP, 0 ¼ else). d These socio-demographic control variables of age, gender, education, and class are measured as in Table 1. Source: CIS2382 and 2384 (pre and post 2000 elections survey).

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The key assumption here, often neglected by practicing researchers, is that the endogenous independent variable – I0 – must be truly exogenous, or it will be a poor instrument. [Refer to the discussion in Wooldridge (2006: 525–540).] In particular, the alleged exogenous variables from the surveys (the Xs) employed to build the instrument must not be causally dependent on available variables (i.e., the causal arrows must run away from the Xs, not to them). Also, these X variables may not be correlated with the model error term. Such conditions are easily met by the usual socio-economic variables. For instance, the variables we use here, such as gender or education, are largely fixed voter conditions. But, the exogenous status of some of the more attitudinal variables may be questioned. Take the variables of parents’ ideology and PP support. Each measures events from sometime in the past, especially parents’ ideology. Still, it might be argued that 1996 PP support is represented in et 1 and is correlated with et. To a lesser extent, the same argument might be applied to parents’ ideology. Finally, it is possible that both these variables are partially dependent on political ideology itself, since each is measured by respondent recall. If I0 is not fully exogenous, after all, then its heavy coefficient is open to the charge of upward bias. To avoid that charge, we impose ever more demanding standards of exogeneity on its construction, in a manner first developed by Be´langer et al. (2006). Therefore, the ideology instrument, I0 , is constructed on the basis of ‘‘broad,’’, ‘‘moderate’’ and, finally, ‘‘strict’’ criteria. The broad condition, a la Fiorina (1981), employs all the socio-economic variables (class, religion, age, gender, education) and all the attitudinal variables (parents’ ideology, 1996 PP support, and consumption of political information). The moderate condition employs all of the foregoing, except 1996 PP support. The strict condition employs only the socioeconomic variables. The coefficients for political ideology and economic evaluation under the broad condition are reported in column 1, Table 2a, and have already been discussed. The coefficients under the moderate condition appear in column 2, Table 2a. What these coefficients show is a serious rise in the economics coefficient (from .55 to .87), and a serious decline in the ideology coefficient (from 3.02 to 2.3). Estimates under the strict condition appear in column 3, Table 2a. Here the ideology coefficient drops still further (from 2.30 to 1.77). Overall, the ratio of effects, ideology to economy, is reduced to just over 2.0 (i.e., 1.77/ .87 ¼ 2.03). It appears, then, that the more rigorously the ideology instrument is constructed, the more diminished is its effect. In contrast, the economics effect rises. These relative improvements in the economic effects are seen even more clearly in Table 2b, where we take up the logic developed earlier (in the discussion of the results from Column 3, Table 1). The notion is that of lagging ideology, but not economics, in order to reduce the contamination from measuring them at the same moment. Further, if ideology is as stable as supposed, this makes additional theoretical sense. When this is done, we see that the economic coefficient increases steadily, as one strengthens exogeneity conditions, while the ideology coefficient

decreases. Under strict exogeneity conditions, the ideologyeconomics ratio hardly exceeds 1.0 (i.e., 1.72/1.28 ¼ 1.34). By these measures, the impact of economics appears still greater. To this point, the question of endogeneity has been confined to political ideology. Nevertheless, it seems quite possible that the original economics evaluations are also endogenous. Specifically, ideology itself might influence economics, as has been strongly claimed for the Spanish case. Therefore, we need to exogenize these sociotropic economic evaluations. There is not as much methodological flexibility here as we would like, because of a relative scarcity of available exogenous variables. Moreover, we cannot employ exactly the same exogenous variables as with I0 , or we would risk a serious collinearity problem. Again, in this exercise we look to previous research about the determinants of electors’ subjective evaluations of the national economy. This literature suggests that individual judgments about the economy contain subjective sources of systematic variation such as their level of information, education, personal experience, social status and/or situation in the labour market (Duch et al., 2000). Therefore we employ the following variables: class, age, gender, education, income, consumption of political information and, following Fiorina (1981, 229–230), personal finances (measured in the posterior time period to secure exogeneity). (Details of the coding for these variables appear at the bottom of Table 3. Clearly, it could be useful to

Table 3 Exogenous Ideology, Exogenous Economic Perception, and the Vote: Panel Tests (Two-Stage, Multinomial Logit Estimates).a Vote for PP versus PSOE (1)

(2)

(3)

Economy ‘ t  1b 1.36 (.21) ** 1.66 (.18) ** 1.68 (.16) ** Ideology’ t  1 (broad x)c 3.10 (.12) ** Ideology’ t  1 (mod x) 2.31 (.12) ** Ideology’ t  1 (strict x) 1.78 (.14) ** Controls (not shown)d Intercepts (not shown) Pseudo-R-sq. .20 .10 .07 N

4081

4081

4081

** ¼ Statistical significance, .01; * ¼ Statistical significance, .05; one-tailed tests. Figures in parentheses are standard errors. a Dependent variable ¼ 1 Vote for the incumbent (PP), 2 Vote for PSOE, 3 Vote for IU, 4 Else, 5 Abstention; Only the coefficients among the two main parties are shown. The complete multinomial results (for all the categories of the dependent variable) are available on request. b Economy’ t  1 ¼ an instrumental variable for national economic evaluation in the pre-election questionnaire, constructed from the following variables: Objective social class (2. Service class, 3 White collar, 4 Skilled manual worker, 5 Unskilled manual worker) where self employed is the category of reference; Personal post-election economic evaluations (from 1-very bad to 5-very good); Income (1 Lower, 2 Middle, 3 Upper, 4 No response): middle is the category of reference; Age ¼ in years; Gender (1 for male and 0 for female); Education (from 1-lowest to 6-highest); Consumption of political information in newspaper (from 1-every single day to 6- never) c the ideology instruments are constructed as in Table 2a,b. d the socio-demographic control variables are measured the same as in Table 2a,b. Source: CIS2382 and 2384 (pre and post 2000 elections survey).

M. Fraile, M.S. Lewis-Beck / Electoral Studies 29 (2010) 210–220

add additional exogenous variables to construct this instrument, but no more candidates appear in the dataset). Again the results of the estimation of this instrumental variable, E0 , are available at: http://www.ipp.csic.es/ Pi-esp/fraile.html [Under a still more strict definition of exogeneity for construction of the economics variable, its coefficient remains strong and significant. In other words, the result reported here does not appear arbitrary. Details of this test are reported in the footnote.]3 One critical question is whether this economic instrument is as ‘‘good’’ an instrumental variable, relatively speaking, as is the strictly defined instrument for ideology. An answer comes from correlating the original variables with their instruments, and comparing magnitudes. For the correlation of I and I0 (strict), r ¼ .35. For the correlation of E and E0 , r ¼ .37. Since these correlations are virtually the same, they stand as equally sound instruments. Furthermore, a series of Hausman exogeneity tests demonstrates that these instrumental variables for ideology and economics are statistically superior, in terms of desirable estimator properties, to the original variables for ideology and economics. (Details of these tests appear in the footnote.) 4 In Table 3, the fully dynamic model (of Table 2a) is reestimated, but this time with these statistically more secure instrumental measures of both ideology and economics. These coefficients, compared to those in previous tables, should be generally preferred, as they are based on more desirable estimator properties. Observe that the economics coefficient is highly significant and fairly stable, across the different exogeneity conditions. In particular, note the last column, where the instruments stand as equally good exogenous variables – the effects of economics and ideology are virtually identical (at 1.68 and 1.78, respectively). In sum, for Spanish voters, how they

3 As an experiment, we dropped personal finances from the construction of the economics instrument, relying strictly on objective social class, income, age, gender, and political information consumption in a replication (of column 2, Table 3). The coefficient of this strict economics instrument stands strong at 1.47, and is highly significant. Nevertheless, we do not favor this instrument over that of original Table 3, because it is a relatively poor instrument in comparison. 4 The Hausman test analyses the exogenous status of an independent variable, through examination of the classical assumption that it is uncorrelated with the error term of the equation. [For a good explication, see Kmenta (1997: 365–366).] Upon application of the test, if statistical significance is found, the conclusion is that the independent variable in question is not exogenous. (Regress endogenous independent variable I on the instruments, and save the residuals, labeling them variable e1*. Do the same for endogenous independent variable E, to obtain variable e2*. Now regress the vote variable on original variables I and E, the controls, and e1* and e2*.) The test under the strict condition (column 3, Table 3) is most demanding, and also shows the endogeneity problem most clearly. Applying a joint significance test on the obtained coefficients of e1* and e2*, the chi-squared ¼ 156.00, significance > 0001. Thus, the original independent variables of ideology and economics are not exogenous, when used in an uncorrected estimation. Further, the same test on the coefficients individually yields, for the coefficient of e1* a chi-squared of 103.11, significance > .0001; for the coefficient of e2* a chi-square of 50. 46, significance > .0001. This implies, further, that the endogeneity problem is more severe for the original ideology variable than for the original economics variables. Interestingly, this conclusion is the opposite of the revisionist argument.

217

evaluated the national economy importantly determined their vote. Indeed, the strength of that evaluation rivals the impact of political ideology itself. 6. Alternatives to instrumental variables estimation Before finally accepting the claimed importance for the economic vote in this contest, it is useful to entertain a critique of its instrumental variables foundation. While we have mapped and justified its steps, virtually any instrumental variables estimates are vulnerable to the charge they rest on arbitrary selection of the exogenous variables needed for their construction. As a leading econometrician noted sometime ago, ‘‘The set of instrumental variables which we may choose is not unique.’’ (Koutsoyiannis, 1984, 383). Because of this condition, we examined alternative instrumental variable sets, with a special eye to fulfillment of the exogeneity standard. (See again Tables 2a,b and 3). Now suppose, in response to these alternatives, we follow a radical abandonment of the instrumental variables approach. Recall the essential question concerns the impact of national economic perception on the vote, uncontaminated by the influence of political ideology. A simple experiment suggests itself. Examine the relationship between the national economic perception variable, E, and declared vote, D, for those voters who have no ideology. If it is true that national economic perception is determined by political ideology, as the revisionists claim, then the observed relationship between E and D should not be statistically significant. That is, economic perception, bereft of its ideological driver, would essentially relate randomly to vote choice. Carrying out this experiment below, we find that this is far from the case. First, we relate economic evaluations, E, at time t (2 ¼ very bad to þ2 ¼ very good) to vote, D, at time t (PP, PSOE, IU, Other, Abstention), within and across ideological groups at time t. Given the categorical dependent variable, Cramer’s V, a nominal measure of association, is preferred. Taking all ideological groups together (2 ¼ extreme left to þ 2 ¼ extreme right), relating E to D yields a statistically significant V ¼ .20 (N ¼ 4343). [All the reported V in this section are statistically significant at .001 or beyond]. If the revisionist argument is correct, this association should plummet when restricted to those without any ideology (i.e., non-respondents on the ideology item). Instead, it hardly moves, registering a statistically significant V ¼ .17. The result repeats itself if those in the Center (score of 0) are judged to be without ideology (i.e., certainly they are without ideology in the Left v. Right sense). Within this Center group, V ¼ .16 (statistically significant). Thus, under either measure of ‘‘non-ideology,’’ economic perception influences the vote about the same as it does in the more general, ideological, population. Additionally, this ‘‘nonideological group’’ (i.e., non-respondent or Center respondent) composes an N ¼ 1602, or 37 percent of the total sample. Note, finally, that the association within the Left (scored 2, 1) and the Right (scored þ1, þ2) ideological groups does not vary much from that of the Center:

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for the Right, V ¼ .13, for the Left, V ¼ .09 (both statistically significant). Overall, then, by these simple tests, the relationship between economic perception and the vote seems little distorted by ideology. Of course, these results, interesting as they are, are based on the simple examination of measures of association while controlling for a third variable. They immediately pose the more causal, multivariate questions asked in the early part of this article. Still, readers, if they wish, can avoid the instrumental variable approach. Recall that Table 1 offers panel models of the vote choice, with extensive control variables and dynamic specifications (but not instruments). There, one still observes relatively strong economic voting effects, ranging from about one-half that of ideology, to almost the equal of ideology, depending on the specification selected. In sum, one may eschew instrumental variables altogether, and still arrive at a robust economic voter effect for this election.

7. Was the 2000 election exceptional? One empirical question is whether the strong economic voting effects observed in the 2000 are exceptional. If, in fact, economic voting appears weak to non-existent in other contexts, then the general argument for an economic vote in Spain would be severely undercut. Unfortunately, as mentioned, we do not have other panel surveys from Spanish elections available for study. Until now, the exclusive form of data analysis for the Spanish case has been cross-sectional. However, we have learned, from current panel investigations elsewhere, that crosssectional economic voting studies potentially underestimate the economic vote (Lewis-Beck et al., 2008). Hence, if we can find a pattern of statistically and substantively significant economic effects in relevant cross-sectional investigations of past Spanish general elections, we might infer something about whether our 2000 results are atypical. Fraile (2005, 224), using logit analysis, estimated a Spanish voting model with a specification similar to ours, for each of the national elections – 1979, 1982, 1986, 1989, 1993, 1996.5 While the sociotropic retrospective economic variable of her model has a somewhat different wording from ours, it is nevertheless useful to examine the magnitude and time trend of these statistically significant coefficients across these six elections from 1979 to 1996: .71, 1.4, 1.9, 2.0, 2.3, 2.1. First, the series gives an enduring presence to economic voting in the Spanish system, since they are always statistically significant (at .05 or higher). Secondly, one observes a tendency for the magnitude of that effect to rise, election after election, suggesting that Spanish voters are ‘‘learning’’ to hold the incumbent responsible for the economy, as they acquire more and more experience in

5 The specification holds vote (1 ¼ incumbent party, 0 ¼ other) to be a function of ideology, age, education, work status, and sociotropic economic evaluations, both prospective and retrospective (Fraile, 2005, 220). The latter measures whether respondents think that economic policies are good or bad (Fraile, 2005, 75).

the electoral game played in this relatively new democracy. [On the basis of a simulation, these economic effects do not appear small; the median probability of an incumbent vote increase is 27%, for those who see the economic policy as ‘‘good or very good,’’ instead of ‘‘bad or very bad.’’ (Fraile, 2005: 148).] Looking at the coefficient pattern from the last three elections, we might expect the logit estimate to be just a bit higher than ‘‘2.0’’ for the 2000 election. Indeed, in a different article, another logit estimation by Fraile (2002, Table 2) suggests it might be close to 2.1. Finally, the higher values for the economic coefficient as the series lengthens may help resolve a sometime aggregate-level paradox, namely an incumbent (e.g., the PSOE elected in 1982) stays in office in the face of bad economic times nationally. Our purpose here is not to resurrect the cross-sectional approach, which we have clearly challenged from the beginning. Rather, we want to place these 2000 economic voting results in temporal context. Clearly, economic voting appears an important presence at least since the 1990s. The fact that it may be larger in 2000 accords with the time trend, and with the application of exogeneity measures in a panel methodology, which tends to make economic effects clearer. A second question is a theoretical one. The prevailing claim that economic voting is trivial, even non-existent, in Spanish elections, raises an important question about the functioning of its democracy. If ideological thinking trumps all other considerations, then the political class has pretty much a free hand in the economic policy arena. The only constraint would be a very broad, longterm one, touching on the borders of capitalism and socialism, respectively. With respect to day-to-day decisions affecting the economy, it could pursue its immediate desires, even reward its own elites, or its own interests. The analysis here helps falsify this claim. Those electors who saw economic deterioration voted against the ruling PP, while those who saw economic progress voted for it. In other words, they responded very much as a reward-punishment theory would predict. This is not surprising, at least if one accepts the retrospective voting idea. Voters do not need to be experts on the economic policies of the government. All they have to know is how those policies are turning out. As Fiorina (1981, 5), the leading exponent of retrospective voting, puts it, ‘‘they know what life has been like during the incumbent’s administration.’’ That is, observing economic conditions all around them, they arrive at conclusions about how things have been, and hold the government accountable on election day.

8. Conclusions In other leading democracies, individual-level survey studies of elections routinely find that economic evaluations are a key determinant of the vote choice. However, that view has recently been challenged by revisionists, who argue that economic evaluations are essentially projections of partisan feelings. No where is this challenge

M. Fraile, M.S. Lewis-Beck / Electoral Studies 29 (2010) 210–220

stronger than from Spain, where local orthodoxy holds that there is no economic voting, because of the reliance of the electorate on left-right ideological thinking. Thus, the Spanish case appears to pose a serious anomaly in the literature. Further, if ideology is the decisive electoral force there, then the normative problem of democratic accountability presents itself. Focusing on the 2000 general election, we do find strong ideological effects, as has been previously argued. But, perhaps surprisingly, we also find significant and sharp economic effects on the vote choice, regardless of the varying test conditions. The simple panel analysis, taking full advantage of the causal assumption of temporality, shows an ideology-economy ratio of the MNL coefficients somewhat over 2:1. (Recall column 2, Table 1). The complex panel analysis with the same specification, but employing the strictest set of exogenity conditions, shows an ideology-economy ratio of about 1:1. (Recall column 3, Table 3). Either of these specifications, the simple or the complex, gives a substantial role to economics. Are these effects too large to be believed? Not necessarily. They are quite in line with simple and complex panel analysis results reported from general elections in the United States, the United Kingdom, and Canada (Lewis-Beck et al., 2008). Political ideology is certainly a dominant, not to say the dominant, long-term force acting on the political behavior of individual Spanish voters. These data confirm that. But they do not confirm that ideology is even close to the exclusive force. In the 2000 general election voters had the opportunity to cast their vote on the basis of government economic performance, and they did so. To the extent this pattern is repeated in other elections, Spanish voters manage to hold their elected officials accountable, as democracies say they should. All and all, the Spanish case seems to have ceased to be an economic voting anomaly in the Western democracies, even assuming it once was. A final cautionary remark is in order. Our conclusions rest on analysis from one election, that of 2000. That is of necessity, since it is only for that election that the requisite panel data are available. Still, this leaves the charge that the 2000 election was atypical, given the economic prosperity of the time, and the unusual leadership difficulties of the left. In response to that charge, one answer lies in our report of the over time crosssectional comparison of economic coefficients from other studies. The economic effects we relay are not undermined by these prior studies. If anything, they are reinforced. Furthermore, we have begun our own pooled (cross-sectional time series) analysis of eight available Spanish national election surveys, which again suggests a broad and strong economic voting effect, regardless of whether the economic indicators are subjective or objective (Fraile and Lewis-Beck, 2010). Nevertheless, more survey panel analysis, of the sort done here remains, to further confirm (or disconfirm) the general economic voting hypothesis for the Spanish case. Fortunately, a usable panel study was conducted for the 2008 elections. Now that it is available for analysis, further testing can take place.

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Appendix. Table 1 The effect of ideology and economy on the vote under independent variable values: probabilities of voting for the two main parties.

Equation (1) (Table 1) Baseline probability Ideology: 2 þ2 Economy: 2 þ2 Equation (2) (Table 1) Baseline probability Ideology: 2 þ2 Economy: 2 þ2 Equation (3) (Table 1) Baseline probability Ideology: 2 þ2 Economy: 2 þ2 Equation (1) (Table 3) Baseline probability Ideology: 2 þ2 Economy: 2 þ2 Equation (2) (Table 3) Baseline probability Ideology: 2 þ2 Economy: 2 þ2 Equation (3) (Table 3) Baseline probability Ideology: 2 þ2 Economy: 2 þ2

Vote for PP

Vote for PSOE

.38

.21

.01 .91

.58 .01

.05 .74

.29 .09

.38

.24

.04 .85

.52 .02

.12 .66

.33 .13

.38

.23

.05 .84

.51 .03

.04 .77

.38 .08

.40

.20

.02 .96

.50 .01

.14 .68

.29 .10

.39

.23

.05 .88

.52 .02

.10 .73

.35 .09

.40

.24

.16 .79

.34 .08

.10 .73

.37 .10

Source: Our elaboration on results of Tables 1 and 3.

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