Abstract
We argue that two different sets of considerations shape the decision to vote or abstain in an election–ethical and non-ethical. First the citizen may vote out of a sense of duty. Failing that, she may vote because she has strong preferences about the outcome of the election. Abstention occurs when neither duty nor a sufficiently strong preference is present. The implication is that while duty and preference each have strong positive effects on turnout, they also have a negative interaction effect, since the impact of preference is much weaker among those with a sense of duty. We present a wide array of empirical evidence that systematically supports our claim that the turnout decision is importantly shaped by this causal heterogeneity. Thus a turnout model misses something fundamental if it does not take into account the effect of civic duty.
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Voting may have a habitual component (Plutzer 2002; Aldrich et al. 2011), but that does not affect our argument. Even in the colloquial sense of “habit” (a powerful addiction to heroin), no one would find it satisfying to hear that an addict injected heroin because he had a habit of doing so. One needs to know the causes of the habit.
One academic economist in a highly ranked department said to us, “If I were a second-rate economist, I wouldn’t vote.” He actually used a more colorful synonym for “second-rate.”.
As indicated above, we believe that instrumental considerations do not play a major role, since on instrumental grounds no one should vote unless turnout is expected to be very low. This does not mean that they play no role at all: for instance, turnout seems to decrease slightly when it rains (Gomez et al. 2007; but see Persson et al. 2014, who find no effect in Sweden). Our claim is rather that expressive and ethical considerations are much more important than instrumental ones.
These results come from the 2004 General Social Survey. The question is: “There are different opinions as to what it takes to be a good citizen. As far as you are concerned personally, on a scale of 1–7, where 1 is not at all important and 7 is very important, how important is it to….”.
We are considering only the literature that examines the decision to vote or not to vote at the individual level. There is a separate stream of research that deals with aggregate variations in turnout across countries or over time. (For a review, see Blais 2006).
There is a substantial literature examining the impact of social pressure on turnout (see especially Gerber et al. 2008), which assumes the presence of a duty norm. The focus of these studies, however, is on how social sanctions contribute to enforcing the norm. Our focus is on how the internalized norm affects voting, independent of pressure.
It is measured through agree/disagree statements that “it is every citizen’s duty to vote in an election” and that “democracy only works properly if most people vote.”.
In an earlier version of this paper, we reviewed what contemporary philosophers and formal theorists have said about the duty to vote, and we distinguished deontological duty from concepts with which it is sometimes confused, such as altruism or conformity to a social norm. To save space, that discussion is omitted here but is available from the authors on request.
All the analyses presented in this paper deal with national, high salience elections. However, our argument applies equally well to low salience elections, the only difference being that both preferences and duty are likely to be weaker for lower-level offices. See Galais and Blais (2016b) for a comparative analysis of the role of duty in different types of elections.
The format is also preferable to the agree/disagree statement format that is usually utilized to measure civic duty. The latter format has been shown to be plagued with acquiescence bias (Schuman and Presser 1981, pp. 202–230).
The switching percentages were higher in the 1956–1960 ANES sample, with corresponding switching rates of 11% and 6%, respectively. We are not certain what accounts for the difference, though sampling error may be the culprit: The early ANES surveys were less good than the later ones, and in addition, each percentage is based on a sample of about 100 turnout switchers. All these percentages refer to the most theoretically relevant question in the ANES battery, which asked whether a person who does not care about the outcome should vote in an election. The other ANES duty items were asked only during the 1956–1960 panel, and their switching percentages are similar.
Our test-retest interval is a year while Converse's was two years. However, as our results show, once the length of the interval between surveys exceeds two or three months, the correlations are little affected by the time period between them. Converse (1964) similarly found that correlations over a four-year period were nearly identical to those over a two-year period.
This specification also allows for those cases in which the citizen actually does have an internalized sense of duty but falls ill or has more important duties on Election Day.
There are a few studies in which interaction effects are considered. Kittilson and Anderson (2011) examine how the impact of political efficacy depends on contextual factors such as the number of parties and party polarization. Soderlund et al. (2011) show that political interest is more closely connected to turnout in second-order elections. Solt (2008) demonstrates that the impact of inequality on electoral participation is strongest among the poor. One study explicitly looks at how sense of civic duty interacts with interest in politics (Blais and Labbé St-Vincent 2011). These authors do not provide, however, a theoretical justification for this interaction, since they are more concerned with examining how political attitudes mediate the effect of personality traits.
The negative sign is not a ceiling effect: As has long been understood, probit and logit models eliminate floors and ceilings (Winship and Mare 1984, 514). Put another way, it is statistically impossible to generate a negative interaction term in probit or logit unless a substantive interaction exists. However, floor and ceiling effects can occur in linear probability models and other approaches that model probabilities directly.
A few weights were very large, occasionally as much as 13 (with the mean weight being close to 1.0). We tried truncating the weights at 3.0 and at 5.0, accepting the biases in hope of better standard errors. We also tried eliminating weighting entirely. The result was too few nonvoters, generally unchanged coefficients apart from the intercept term, no changes in substantive findings, and only slightly better standard errors. We therefore adhere to the original CCAP weights throughout this paper.
To avoid too many near-empty cells in these tables, we have collapsed the bottom two categories of caring about the election and the middle two categories of duty. In the probit specifications, however, we have kept the original categories, since they preserve more information and, as expected, provide a somewhat better fit.
We verified the existence of a probit-scale interaction effect in a more robust way. If the sample is divided into subsamples with various levels of Duty (combining as necessary to get minimal sample sizes), and if Preference and Duty interact, then when the model is estimated separately in each subsample, the probit coefficient of Preference should fall across the subsamples as Duty increases. The same is true for the effect of Duty when the sample is subsetted by values of Preference. They both do. Thus the negative interaction term is the data talking, not just the model.
We report the usual two-sided tests to accord with convention. However, they make little sense here. A statistically significant negative coefficient for duty or care, for example, would be interpreted as bad sampling luck or mismeasurement, not as a substantive finding. Hence in the tables, a coefficient marked significant at 10% is actually significant at 5% on a more meaningful, one-sided test.
Kam and Franzese (2007, Chaps. 3, 5, and Appendix B) is a clear recent exposition of how to compute the underlying partial derivatives and their sampling errors in graphs like Fig. 2. However, unlike them, we have chosen to represent marginal effects on the probit scale in Fig. 2 rather than on the probability scale, since only the probit-scale effects are stable across samples when the theoretical model is stable. (See also Brambor et al. 2006 and Berry et al. 2009.).
Interest is a three-point scale based on the response to a question about the respondent’s level of interest in politics and current events. (“Not sure” was combined with “not much.”) Strength of party identification is a four-point scale, from strong to weak to leaning to Independent, based on the classic Michigan item. Age range runs from 18 to 90. Education is the five-point scale provided in the survey. In all cases, skipped or not asked questions were coded as missing. We used interest in October to give it every chance to pick up endogenous short-term campaign effects that might be correlated with duty, though interest measured in January is more meaningful theoretically and actually has a larger coefficient. However, either choice has only the tiniest effects on the duty and preference coefficients.
The survey also had a 12-item battery of political information questions. (“Skipped” was coded as “don’t know.”) The percent answered correctly had only a small, statistically insignificant effect on turnout, and it left the coefficients on duty, preference, and their interaction virtually unchanged.
In this and in many other specifications, the impact of being African-American was relatively small, statistically insignificant, and had a negative sign in an election in which black turnout rose substantially, which gives evidence that our key variables are capturing its effect.
At our suggestion, the 2012 ANES survey adopted a question wording for duty very near our own. (They retained the old ANES dichotomous item for caring about the election.).
There are too few contributors in the 2008 CCAP to allow the same test to be made with those data. All but one of the donors were in the top category of preference, so that the model was very nearly unidentified, standard errors exploded, and a considerable number of observations were completely determined. We also could not test talking about politics in the 2012 ANES since its only similar measure asked the respondents whether they had talked to anyone about voting for or against a candidate or party, a measure of proselytization possibly related to duty and quite different from a pure measure of engagement.
The face-to-face sample had very few respondents choosing intermediate values for duty, so that it became essentially dichotomous. (The Internet sample was better.) The ANES preference variable is dichotomous by design. The resulting large amount of measurement error in both variables attenuates the coefficients, as it also does in the ANES validated vote sample in column 2, where the duty variable was also dichotomous by design.
We took the data from the ANES cumulative file, whose codebook recommends against weighting. These ANES surveys also allowed us to test another counter-hypothesis to our argument, namely that duty appears to predict (reported) turnout because those with a sense of duty are more likely to say that they turned out when they did not. However, no such difference appears in these data.
Here, interest is interest in federal politics as measured in the second wave, age is scaled from 18 to 90, preference is caring which party wins, and duty is duty to vote in federal elections. In other respects, the care and duty questions are worded the same as in the 2008 CCAP. The survey weights supplied by the vendor are used.
Throughout our analyses, we have argued that the impact of preference is stronger among those with a weak sense of duty. Another interpretation is that the impact of duty is stronger among those with weak preferences. In this paper, we focus on duty because this is our main contribution, but preference and duty play equally important roles in our model.
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Acknowledgements
We thank Charles Beitz, Robert George, and Alan Patten for helping us navigate the political theory literature on political obligation and the duty to vote. Larry Bartels and Aram Hur gave us the benefit of a careful reading of an early version. The Princeton Department of Politics supplied research funds, as did the Social Sciences and Humanities Research Council of Canada. We also benefited from dedicated and professional research assistance by Simon St-Vincent and Jason Roy. We thank the reviewers for their very helpful comments and suggestions. The data for replication are posted at openICPSR.
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Blais, A., Achen, C.H. Civic Duty and Voter Turnout. Polit Behav 41, 473–497 (2019). https://doi.org/10.1007/s11109-018-9459-3
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DOI: https://doi.org/10.1007/s11109-018-9459-3