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Political Behavior

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28.03.2016 | Original Paper

Unresponsive and Unpersuaded: The Unintended Consequences of a Voter Persuasion Effort

verfasst von: Michael A. Bailey, Daniel J. Hopkins, Todd Rogers

Erschienen in: Political Behavior | Ausgabe 3/2016

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Abstract

To date, field experiments on campaign tactics have focused overwhelmingly on mobilization and voter turnout, with far more limited attention to persuasion and vote choice. In this paper, we analyze a field experiment with 56,000 Wisconsin voters designed to measure the persuasive effects of canvassing, phone calls, and mailings during the 2008 presidential election. Focusing on the canvassing treatment, we find that persuasive appeals had two unintended consequences. First, they reduced responsiveness to a follow-up survey among infrequent voters, a substantively meaningful behavioral response that has the potential to induce bias in estimates of persuasion effects as well. Second, the persuasive appeals possibly reduced candidate support and almost certainly did not increase it. This counterintuitive finding is reinforced by multiple statistical methods and suggests that contact by a political campaign may engender a backlash.

Vorheriger Artikel National Party Division and Divisive State Primaries in U.S. Presidential Elections, 1948–2012

Nächster Artikel Negative Advertising and the Dynamics of Candidate Support

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The data set and replication code are posted online at https://dataverse.harvard.edu/dataverse/DJHopkins. Due to their proprietary nature, two variables employed in our analyses are omitted from the data set: the Democratic performance in a precinct and each respondent’s probability of voting for the Democratic candidate.

Strategies to study persuasion include natural experiments based on the uneven mapping of television markets to swing states (Simon and Stern 1955; Huber and Arceneaux 2007) or the timing of campaign events (Ladd and Lenz 2009). Other studies use precinct-level randomization (e.g. Arceneaux 2005; Panagopoulos and Green 2008; Rogers and Middleton 2015) or discontinuities in campaigns’ targeting formulae (e.g. Gerber et al. 2011).

In a related vein, Shi (2015) finds that postcards exposing voters to a dissonant argument on same-sex marriage reduce subsequent voter turnout.

Experimental studies also rely on self-reported vote choice, not the actual vote cast. This is less of a concern, as pre-election public opinion surveys like this one typically provide accurate measures of vote choice (Hopkins 2009).

Such support scores are commonly employed by campaigns. To generate them, data vendors fit a model to data where candidate support is observed, typically survey data. They then use the model, alongside known demographic and geographic characteristics, to estimate each voter’s probability of supporting a given candidate in a much broader sample. The specific model employed is proprietary and unknown to the researchers. The Pearson’s correlation with a separate measure of precinct-level prior Democratic support is 0.47, indicating the importance of precinct-level measures in its calculation in this data set. For more on the use of such data and scores within political science, see Ansolabehere et al. (2011), Ansolabehere et al. (2012), Rogers and Aida (2014) and Hersh (2015).

This age skew reduces one empirical concern, which is that voters under the age of 26 have truncated vote histories. Only 2.1% of targeted voters were under 26 in 2008, and thus under 18 in 2000.

Specifically, voters were coded as “strong Obama,” “lean Obama,” “undecided,” “lean McCain,” and “strong McCain.”

We can do additional analyses to approximate the effect of the treatment on people who actually spoke to the canvassers (the so-called Complier Average Causal Effect; see Angrist et al. 1996), and report the results in the Conclusion.

For the full regression, see the first column of Table 6 in the Appendix.

Similar results for the phone and mail treatments show no significant differences across groups.

For the corresponding regression model, see the second column of Table 6 in the Appendix.

Voters under the age of 26 would not have been eligible to vote in some of the prior elections, and might be disproportionately represented among the low-turnout groups. We have age data only for 39, 187 individuals in the sample. The negative effects of canvassing in the zero-turnout group persist (with a larger confidence interval) when the data set is restricted to citizens known to be older than 26.

The effects for phone calls are generally similar, but not statistically significant (see Table 9 in the Appendix).

For example, Enos et al. (2014) find that direct mail, phone calls, and canvassing had small effects on turnout for voters with low probabilities of voting, large effects for voters with middle-to-high probabilities of voting, and smaller but still positive effects for those with the highest probabilities of voting.

Results using logistic regression are highly similar.

In separate, ongoing research, we use the turnout results described above as a benchmark with which to evaluate each of these methods.

As Little et al. (2012) explain, “weighted estimating equations and multiple-imputation models have an advantage in that they can be used to incorporate auxiliary information about the missing data into the final analysis, and they give standard errors and p values that incorporate missing-data uncertainty”(1359).

But that fact also means that the “implied joint distributions may not exist theoretically”(Buuren et al. 2006, p. 1051). Still, that important theoretical limitation does not prevent MICE from working well in practice (Buuren et al. 2006).

To examine the performance of our model for multiple imputation, we performed tests in which we deliberately deleted 500 known survey responses from the fully observed data set (n = 12,442) and then assessed the performance of our imputation model for those 500 cases where we know the correct answer. In each case, we used the full multiple imputation model to generate five imputed data sets for each new data set, and then calculated the share of deleted responses which we correctly imputed. The median out-of-sample accuracy across the resulting data sets was 74.9 %, with a minimum of 73.3 % and a maximum of 76.0 %. This performance is certainly better than chance alone.

In fact, the associated p value is less than 0.002, meaning that the finding would remain significant even after a Bonferroni correction for multiple comparisons to account for the analyses of the phone and mail treatments.

The associated 95 % confidence interval spans from −3.03 to −0.60.

We could add additional covariates that only affect this equation without affecting our discussion below. The existence of such variables is commonly necessary for empirical estimation of selection models, although it is not strictly required, as these models can be identified solely with parametric assumptions about error terms.

Throughout these analyses, we drop our measure of respondents’ age, which is the only independent variable with significant missingness.

Here, \(\delta\) is set to 0.0001.

Siddique and Belin (2008a) report that a value of \(k=3\) works well in their substantive application, while Siddique and Belin (2008b) recommend values between 1 and 2.

Still, even in light of this potential to under-estimate variance, Demirtas et al. (2007) demonstrate that the small-sample properties of the original ABB are superior when compared to would-be corrections.

IPW requires data that are fully observed with the exception of the missing outcome. We thus set aside 20 respondents who were missing data for covariates other than age or Obama support.

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Titel: Unresponsive and Unpersuaded: The Unintended Consequences of a Voter Persuasion Effort
verfasst von: Michael A. Bailey
Daniel J. Hopkins
Todd Rogers
Publikationsdatum: 28.03.2016
Verlag: Springer US
Erschienen in: Political Behavior / Ausgabe 3/2016
Print ISSN: 0190-9320
Elektronische ISSN: 1573-6687
DOI: https://doi.org/10.1007/s11109-016-9338-8

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