Abstract
Only a small portion of Americans make campaign donations, yet because ambitious politicians need these resources, this group may be particularly important for shaping political outcomes. We investigate the characteristics and motivations of the donorate using a novel dataset that combines administrative records of two types of political participation, contributing and voting, with a rich set of survey variables. These merged observations allow us to examine differences in demographics, validated voting, and ideology across subgroups of the population and to evaluate the motivations of those who donate. We find that in both parties donors are consistently and notably divergent from non-donors to a larger degree than voters are divergent from non-voters. Of great interest, in both parties donors are more ideologically extreme than other partisans, including primary voters. With respect to why individuals contribute, we show that donors appear responsive to their perception of the stakes in the election. We also present evidence that inferences about donor ideology derived from the candidates donors give to may not closely reflect the within-party policy ideology of those donors. Overall, our results suggest that donations are a way for citizens motivated by the perceived stakes of elections to increase their participation beyond solely turning out.
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Notes
This work also relies on self-reported turnout.
See the Supplementary Information (SI) for a discussion of the accuracy of reported donation behavior in our merged data.
This is important because while many party activists are likely donors, we do not know how many donors are activists.
We compare among registrants because almost all donors are registered and because those who are not registered cannot vote.
We evaluate this matching process in the Supplementary Information.
Our comparison of voters to non-voters does not condition on donation behavior. As footnote 10 shows, almost all donors vote.
Analysis excluding leaners is available upon request.
For an account of differences between Democratic and Republican donors, see Francia et al. (2005).
Voters, in this case, are those with a validated turnout record from the 2012 presidential contest.
6.1 percent of donors in these data are not validated to have voted in either the 2012 general election or a 2012 primary election. See SI Table S1 for analysis of participation by donor status.
We apply Stata’s IPF factor command to CCES variables CC320, CC321, CC322_1-CC322-6, CC324, CC325, CC326, CC327, CC328, and CC329. We first break each categorical item into a set of dummy variables for all responses (including missing response) for a single-factor analysis. Factor coefficients are reported in SI Table S2. The first factor has an eigenvalue of 4.3. We exclude the roll call items, CC332A-J, out of concern that they reflect bills congress has already considered. Respondent “votes” on those bills may therefore proxy political sophistication, confounded with donor status, rather than actual differences in policy ideology. We have estimated these models excluding the immigration items out of concern that they are related to foreign policy positions and find highly similar results. Similarly, we have also estimated these measures excluding the binary response items (CC322_1-CC332_6 [immigration policy] and CC326 [gay marriage]). Results are again highly similar (Compare SI Table S3 to Table 2).
We also investigated preferences over tax policy, because one might imagine that wealthy Democratic contributors might oppose higher taxes and therefore prevent more populist Democratic tax policies. However, we found that Democratic contributors of more than $1,000 did not differ in their preferences over tax policy from non-contributors, and that contributors of less than $1,000 preferred tax cuts for the middle class, but not the wealthy, by about 10 percentage points more than non-contributors. For Republicans, support for tax cuts is high across the board, and increasing in size of contributions. Analysis available from the authors on request.
One concern is that random measurement error may be larger for less sophisticated respondents, making them appear more centrist, and sophistication may be correlated with other factors (e.g., income and education) that predict giving. For this reason, we have also replicated our analysis for respondents with at least a 4-year college degree and find similar results with smaller magnitudes. See SI Table S4.
Because most contributors also vote in general and primary elections, and because most primary voters also vote in general elections, we order the comparisons in this way.
These are items CC320, CC321, CC324, CC325, and CC327.
This is consistent with work by Jacobs and Page (2005), who find that the mass public’s attitudes on foreign policy appears largely unrelated to elite foreign policy preferences.
There are clearly other factors that motivate the decision to give beyond those considered here, such as social networks, etc. This analysis captures one factor of this choice.
The other two categories of donors identified by Francia et al. are investors (those who donate for personal material incentives) and intimates (those who donate for social reasons).
Green et al. (2015) find suggestive evidence that policy interests motivate donation behavior.
Of course, an equilibrium in which only extremists contribute and therefore win on policy grounds may not be sustainable if there are centrist voters.
This model abstracts away from the question of where party positions come from or why, ex ante, the parties do not converge. In doing so, it also sets aside the question of whether donations are motivated by a desire to shape primary election outcomes.
An alternative phrasing of the question is “which voters are more likely to make an expressive (rather than instrumental) contribution given the stakes they perceive?” In either case, whether the choice is motivated by a desire to influence the election or just to express one’s view about it, the central intuition is the same: perceived stakes will increase the benefit of contributing.
For simplicity, this exposition ignores the question of whether individuals differ in their assessments of each party’s chances of winning office absent contribution behavior, how contributions influence elections, or the individual cost of contributing.
We have also estimated models in which we assume policy loss is linear in the relative distance between the two parties. For the entire sample, as well as for Democrats and Republicans separately, we continue to find evidence that greater expected policy loss is associated with a greater willingness to contribute.
We obtain similar results if we instead use either the party or candidate placement measures. Results are available upon request.
These measures rely on different scalings of related spatial placement measures. As such, this raises the possibility of collinearity, which will tend to inflate the standard errors of regression estimates and make it hard to find statistically significant results. We present a full correlation matrix for the different spatial measures in SI Table S6.
Angrist and Pischke (2009, p. 103) discuss the relative merits of OLS versus limited dependent variables models and argue that OLS and limited dependent variables models (e.g., Logit) produce very similar point estimates for the marginal effects of explanatory variables. In our case, OLS and logit estimates of the influence of squared distance are very similar.
For a Democrat whose self-placement is liberal and perceives the Democratic Party as very liberal, moving from perceiving the Republican Party as moderate to perceiving it as very conservative increases the predicted rate of contribution by 6.9 points. This is a proportional increase of 46 % relative to the baseline for Democrats in this sample.
Such a pattern could also arise through measurement error, if less engaged individuals give less and also provide more centrist assessments of the party’s positions.
Because our measure of policy ideology is taken in October 2012, we also created a CFscore for each respondent that is specific to the year 2012. Specifically, we calculated a dollar-weighted average CFscore of the candidates to which the donor gave in the 2012 cycle and use this as the donor’s 2012 CFscore for robustness analysis that appears in the SI. We find similar results with this 2012 CFscore.
If instead of conditioning on party we simply divided the CFscores into 3 bins, with one cut at −0.5 and another at 0.5, we see similar results. Within the bottom and top bins, CFscores are only weakly related to variation in individual-level policy ideology (r = 0.14 in both conditions).
We have also replicated this analysis using different outcome measures: Approval for Obama minus Approval for Congress and self-placement ideology. Self-placement ideology helps mitigate concerns about differences in statistical procedures to scale ideology generating part of the discrepancy. For both measures, we see large differences between the parties, but within parties, the CFscore measure does not predict much observed variation. See Figure S12 in the SI for analysis using self-reported ideology.
An alternative explanation for the greater likelihood of donations from more extreme individuals is candidate fundraising behavior (see, e.g., Johnson 2010). Candidates at the fringes of the ideological spectrum may have a harder time raising funds from more pragmatic PAC or corporate donors, and so instead make efforts to reach out to ideological individual candidates.
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We thank Dan Biggers, Adam Bonica, Joshua Clinton, Shigeo Hirano, Gary Jacobson, Lynda Powell, Brian Schaffner, Chris Tausanovitch, Danielle Thompson, the editor, and anonymous reviewers for feedback. A replication archive is available at http://huber.research.yale.edu.
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Hill, S.J., Huber, G.A. Representativeness and Motivations of the Contemporary Donorate: Results from Merged Survey and Administrative Records. Polit Behav 39, 3–29 (2017). https://doi.org/10.1007/s11109-016-9343-y
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DOI: https://doi.org/10.1007/s11109-016-9343-y