Abstract
Our study investigates how and why racial prejudice can fuel white opposition to gun restrictions. Drawing on research across disciplines, we suggest that the language of individual freedom used by the gun rights movement utilizes the same racially meaningful tropes as the rhetoric of the white resistance to black civil rights that developed after WWII and into the 1970s. This indicates that the gun rights narrative is color-coded and evocative of racial resentment. To determine whether racial prejudice depresses white support for gun control, we designed a priming experiment which exposed respondents to pictures of blacks and whites drawn from the IAT. Results show that exposure to the prime suppressed support for gun control compared to the control, conditional upon a respondent’s level of racial resentment. Analyses of ANES data (2004–2013) reaffirm these findings. Racial resentment is a statistically significant and substantively important predictor of white opposition to gun control.
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Introduction
Do racial predispositions influence the gun policy preferences of white Americans? Despite a substantial racial divide in gun control attitudes, with a much larger proportion of whites than blacks opposing gun control measures (Pew Research Center 2013), this question has received scant attention from scholars of racial prejudice. Only two studies have sought to link prejudice to gun policy attitudes. Both studies tested whether racial attitudes, theorized as responses to fear of racialzed crime, influenced whites’ gun policy attitudes. Kleck (1996) produced null results and O’Brien et al. (2013) reported mixed results suggesting that racialized fear of victimization may not be a strong predictor of white gun policy attitudes.
With the theory of racial resentment as our starting point, we revisit the relationship between white gun policy preferences and racial prejudice. We develop a systematic yet historically situated explanation (Mahoney 2000; Pierson 2004) for why a substantial portion of whites oppose further gun restrictions. According to historians, opponents of the civil rights movement used the language of rights and equality to defeat anti-discrimination laws and affirmative action (Self 2003; Sugrue 1995; Hosang 2010). This discourse became central to the politics of resentment of the 1960s and 1970s and formed the basis for a new expression of racial prejudice known as “racial resentment.” As a result, the language of equality and rights has acquired racialized meanings for a portion of the white population. We argue that the contemporary debate over guns reflects the temporal and strategic confluence between the narrative of racial resentment (Kinder and Sanders 1996; Kinder and Sears 1981) and the narrative of the gun rights movement which developed in the 1970s. We suggest that the adoption of this language by the gun rights movement helped ensure that racial prejudice colors whites’ gun policy attitudes.
Our hypothesis that racial predispositions are a key factor that shapes gun policy attitudes is supported by the results of a survey priming experiment and extensive analyses of multiple waves of the ANES. Our work offers a new explanation for the relationship between racial prejudice and gun policy preferences, and we make significant methodological and empirical contributions to the literature. Methodologically, we develop and implement a priming experiment to scrutinize the relevant causal link. Empirically, we expand upon the measures of gun control attitudes and the period of time analyzed using cross-sectional and panel analyses.
The Politics of Racial Resentment and Gun Policy Preferences
Understanding the relationship between racial resentment and gun policy attitudes requires a brief discussion of the politics of resentment and the emergence of racial resentment as a dominant expression of prejudice. By racial resentment, we mean an expression of prejudice that rests on elevating individual rights and traditional values associated with individualism to an ideal status, and castigates blacks as lacking in those values not because of innate characteristics but as a result of a political choice to push for “special” rights from the state (Kinder and Sanders 1996; Sears and Henry 2003). Such beliefs emphasize whites’ alleged moral superiority and ignore the structural advantages of whiteness.
Racial resentment highlights the perceived failure of blacks (but not whites) to live by traditional values of individualism, independence, and hard work. Blacks are viewed as people who depend on government for “special” rights and because of that they constitute a “moral threat” (Kinder and Sanders 1996, 108). This moral threat transcends the individual level and rises to the political because blacks as a group are perceived as willing to trade individual liberty and “equal rights” for group benefits. Although not all whites express feelings of racial resentment, this form of prejudice is more socially acceptable than “old style” forms of prejudice rooted in biological racism that endorse social distance and stereotypes (Sears et al. 2000; Kinder and Sanders 1996; Schuman et al. 1997). Studies show high levels of racial resentment among whites persisting over time (Tuch and Hughes 2011).
The emergence of racial resentment as a dominant form of prejudice is historically contingent. Supporters of racial exclusion have marshalled liberal tropes such as individual rights and the idea of a social contract in support of policies that explicitly or implicitly contradicted these very same liberal values (Stears 2007; King 1999; Dudas 2005). According to Lipset and Raab, classical liberal ideals such as rights and equality have been appropriated to form “the substance of extremist threats to [American] democratic life” (Lipset and Raab 1970, 30).
The modern day politics of resentment has its roots in the post-WWII era. As a fledgling civil rights movement sought to desegregate schools and neighborhoods and end discriminatory employment practices in the public and private sectors, white segregationists responded with violence (Klarman 1994). At the same time as “massive resistance” took place in the South, in other parts of the country white reactionaries developed new arguments in defense of white privilege. Historians have demonstrated that white movements strategically used the language of classical liberalism to create juxtapositions between two types of rights: race-neutral rights assigned to a specific virtuous category of citizens and “special rights” associated with racial and ethnic groups (Hosang 2010; Self 2003; Sugrue 1995). Examples of this juxtaposition include “homeowner rights” used in defense of residential segregation, “taxpayer rights” marshalled against welfare programs and affirmative action, or “victims” rights employed in support of punitive criminal justice policies. In each case, the trope of “rights” was used in defense of white privilege.
White movements viewed government protections against systematic private discrimination as state-sanctioned, preferential treatment that elevated the status of blacks at the expense of whites. These “special rights” were described as un-American and a threat to traditional values of individual merit and equality of opportunity. Dudas argues that “stigmatizing egalitarian politics as subversive and validating defenses of hierarchy as patriotic, activists’ rights talk inflate[d] their resentment; it convince[d] them that their opposition [was] necessary for protecting the American way of life…Special rights talk simultaneously legitimize[d] and motivate[d] opposition to egalitarian social change” (Dudas 2005, 725).
The emergence of the gun rights movement in the 1970s needs to be situated in this historical and political context. Through the 1960s, access to most types of guns was unfettered by federal regulation and despite state and local gun control laws, whites had easy access to firearms (Cottrol and Diamond 1995). In the late 1960s, states responding to the rising political unrest and riots that followed the Civil Rights Movement instituted a variety of gun restrictions. Threatened whites perceived urban violence as either evidence of innate black criminality or as organized and politically motivated (Thernstrom and Thernstrom 1999, 170–171). In 1967, in the wake of the mobilization of the Black Panthers, California passed the Mulford Act which restricted citizens’ ability to carry guns in public spaces (Winkler 2011a). In 1968, Congress introduced national restrictions to gun sales and ownership. Gun laws were tightened again in the early 1990s (e.g., background checks, assault weapons ban, waiting periods).
Through the 1960s, organizations such as the NRA supported gun control laws (Winkler 2011a). However, following a leadership change in 1977, the NRA developed an extreme gun rights position which it defended with a narrative about “(inalienable) gun rights” afforded to the “law abiding citizen.” Gun activists allied themselves with the leadership of the “New Right” movement which had emerged out of the white resistance movements of the 1950s–1960s and the George Wallace political organization (Siegel 2008), and adopted the language of rights and equality to justify an uncompromising gun rights position.
By the 1990s, the dominant gun narrative presented guns not as tools and sporting equipment as it was the case in the 1950 and 1960s (Burbick 2006), but as symbols and instruments of liberty (Melzer 2009; Winkler 2011a). The new narrative was modelled on other seemingly race-neutral rights discourses of the time and played on the same tropes used by prior white movements. Gun owners were cast “as freedom fighters bravely defending all individual rights and freedoms” (Melzer 2009, 65), while firearms and marksmanship signify “morality” and “good citizenship” (Cooper 2001). Guns are “emblem[s] of responsibility, self-sufficiency, personal independence” (Heston 2000, 46).
In labeling as “enemies of freedom” the groups (e.g., minorities and supporters of the rights of the accused) that sought state protection from violence through gun control, the gun rights narrative created yet another juxtaposition of “equal” and “special” rights and became consonant with the themes that undergird racial resentment. Similarly, juxtapositions of “law abiding citizens” and “criminals” were evocative of racialized themes as crime has long been associated with blacks in the white mind (e.g., Mendelberg 2001; Peffley and Hurwitz 2010; Schaller et al. 2003).
As a result of the racialization of the “rights” trope in the post-WWII era, modern day prejudiced whites tend to gravitate toward policy positions presented in race-neutral “rights” terms because these positions are implicitly perceived as advantaging whites. The saturation of the gun advocacy narrative with this language coupled with the specific juxtaposition between “law abiding citizens” and “criminals,” have ensured that prejudiced whites are more likely to support gun “rights” –a position consistent with perceived defense of white privilege- and oppose gun control. This theory does not suggest that all or most white gun owners are prejudiced, or that all or most white people who oppose gun control do so because of racial considerations. We also, do not suggest that racial resentment is the most important driver of opposition to gun control. Our goal here is to show that prejudice is an important and undertheorized factor in a literature that has primarily focused on self-interest or values (e.g., Wolpert and Gimpel 1998; Kleck 1996; Kahan and Braman 2003; Braman and Kahan 2006). Our intention is not to devalue the contribution of these well studied perspectives, but to offer an additional explanation for public opinion on gun policy.
Based on the above theoretical expectations, we hypothesize that:
H1
Whites whose racial considerations are experimentally primed are more likely to oppose gun restrictions than are those in the control group.
H2
Racial resentment, as measured by the symbolic racism scale, is a key predictor of opposition to gun control. Individuals who score high on racial resentment are more likely to oppose gun control than those who score low on racial resentment.
An Alternative Hypothesis: Racialized Crime and Attitudes toward Guns
Research is psychology shows that white Americans associate crime with blacks (e.g., Schaller et al. 2003) and blacks with danger and crime (e.g., Judd et al. 2004; Payne 2001). There is also a significant, though contested, body of research that links fear of crime to gun ownership (for literature review, see: Kleck 1997). A number of early studies in the 1980s viewed white gun ownership as an aggressive response to crime precipitated by racial prejudice. Using attitudinal or contextual measures of racism and racial threat, these studies produced findings that indirectly linked racial factors to firearm ownership (Young 1985; Lizotte and Bordua 1980). However, an early attempt to link racial prejudice to gun policy preferences produced null results (Kleck 1996).Footnote 1
The relationship between fear of black crime and gun attitudes is not easy to deduce. O’Brien et al. (2013) argued that because prejudiced whites conflate blacks with violent crime, they should be more likely to oppose gun control. However, there is not a clear logical basis for the direction of this hypothesis. As gun rights supporters have pointed out, historically, it has been racist whites who supported the disarmament of blacks (Winkler 2011b; Tahmassebi 1991). It is thus equally plausible that prejudiced whites may support gun control in an effort to get guns away from black criminals. Indeed, there is evidence that fear of crime correlates with support for gun control (Celinska 2007; Kahan and Braman 2003). As we will show later, our ANES analyses also support this direction of causation.
The results of the O’Brien, et al. (2013) study suggest that fear of black crime may not be the motivating factor in white attitudes toward gun policy. The study shows no statistically significant relationship between the “blacks violent” stereotype question and gun control attitudes. The AMP, an implicit measure of racism is also not statistically significant. However, they do find stronger, though still mixed, results when using the symbolic racism measure. The fact that a measure that directly relates to racial violence and criminality does not produce results but that the symbolic racism measure is statistically significant provides support for our hypothesis that attitudes towards gun control are related to new forms of racism and identity more so than to instrumental concerns about crime.Footnote 2
A Note on the Measure of Symbolic Racism
Since it was first published, the measure of symbolic racism/racial resentment has generated a number of important criticisms. One serious concern is that it conflates racial attitudes and racial policy preferences (Carmines et al. 2011). Another is that it measures attribution style rather than racial considerations (Gomez and Wilson 2006). The most important and relevant criticism, given our study, is that it is really a measure of political conservatism rather than racial bias (Sniderman et al. 1991; Sniderman and Piazza 1995; Feldman and Huddy 2005). A recent analysis suggests that both racial attitudes and political attitudes are captured by the measure (Neblo 2009).
Symbolic racism has been validated as a measure of racial attitudes in a large number of studies (Kinder and Sanders 1996; Kinder and Mendelberg 1995). Research on emotions also supports the idea that racial resentment is a phenomenon distinct from conservatism and ideology (Banks and Valentino 2012). Furthermore, recent research on racial resentment has expanded, for example using the measure to explain white attitudes toward responses to political campaigns (Mendelberg 2001), the Obama presidency (Tesler and Sears 2010), and healthcare reform (Henderson and Hillygus 2011). Our experiment (as will be discussed) also validates the use of symbolic racism as a measure of racial attitudes as there is no theory to expect that the activation of racial considerations would be contingent on any measure of conservative beliefs.
Priming Experiment
Our survey experiment is designed to do two things: (1) provide causal evidence of a relationship between the activation of racial considerations and whites’ gun control preferences; (2) show that the relationship is moderated by racial resentment as measured through symbolic racism. The experiment was conducted online using a sample of 1205 white respondents from an opt-in online panel. A total of 1179 subjects are included in the analysis.Footnote 3 The study fielded in December 2013, and had a mean completion rate of 27 min.Footnote 4 Quotas on age within gender were used to ensure that the sample reflected these two demographic characteristics of the US white population. Although the demographics and political preferences of the subjects closely approximate those of the white population, the subject pool is a convenience sample.Footnote 5
Under the guise of a cognitive test, we exposed 600 survey participants who self-identified as white to three pictures of the faces of black individuals and another three of white individuals. Respondents were asked to look at the pictures and assess the attractiveness and likeability of the individual. We used the pictures employed in the Implicit Association Test (IAT) which have been designed to activate racial attitudes exclusively (e.g., Greenwald et al. 1998). There is an extensive literature in psychology that uses the IAT pictures or similar primes to activate racial considerations (e.g., Payne 2001; Payne et al. 2002).Footnote 6 Example images are reproduced in Appendix A.Footnote 7
Our prime is supraliminal rather than subliminal: respondents are fully aware of exposure to the prime. Research suggests that supraliminal primes can be effective in activating existing predispositions as long as the respondents are not able to make the connection between the prime task and the items under investigation (Fiske and Taylor 2013). However, there is always the risk that respondents may become aware of the true purpose of the prime and of the study and they may adjust their response accordingly. The short task associated with the prime and its supraliminal nature makes this a relatively weak prime.
Subjects’ attitudes toward guns and gun regulation were assessed based upon responses to 13 specific policy proposals and nine belief statements about guns. In response to the 13 policy proposals, subjects were asked to indicate their level of opposition/support on a 4-point scale. In response to the nine belief statements, subjects were asked to indicate their level of disagreement/agreement on a 4-point scale. An exploratory factor analysis of the 13 policy proposals indicates that seven items fall on the first dimension (eigenvalue = 4.9).Footnote 8 The face validity of all seven items is high. A separate exploratory factor analysis of the nine belief statements indicates that eight items fall on the first dimension (Eigenvalue = 3.6). The wording and descriptive statistics of all fifteen items used in the following analyses are provided in Online Appendices A and B.
We analyze the effect of the treatment on each of the 15 items using OLS.Footnote 9, Footnote 10 The coefficient associated with the treatment is negative in all OLS analyses. Based upon ordered logit analyses, the coefficient associated with the treatment is statistically significant at the p < 0.05 level (one-tailed) in six cases. The six items for which the treatment had a statistically significant direct effect are indicated with an asterisk in Appendix A. The negative coefficients in all 15 cases suggest that exposure to the treatment reduced support for gun regulation. That 37 % of the coefficients are statistically significant at the p < 0.05 level raises the question of why the treatment has a statistically significant direct effect on some items but not others. There are a number of possible explanations. For example, it may be that a stronger prime would have produced results in which a higher percentage of the coefficients were statistically significant. Another possibility is that each of the items taps the underlying attitude imperfectly and to varying degrees.
Assuming that this latter possibility is indeed the case, we construct two additive indices, one based upon the seven policy proposals (alpha = 0.858) and the other based upon the eight belief statements (alpha = 0.859). Summary statistics on the two indices are reported in Appendix B. While both indices are somewhat skewed, OLS would appear to be a reasonable approach to modeling them (Figure A1. In Online Appendix).
Table 1, columns 1 and 4 provide the results of the OLS analyses using the two indices as dependent variables. Both coefficients associated with the treatment are negative and statistically significant at the p < 0.05 level. Subjects who received the treatment had a mean policy index score that was −0.194 units lower than those in the control group, on average, controlling for all other variables in the model. Likewise, subjects who received the treatment had a mean belief index score that was −0.205 units lower than those in the control group, on average, controlling for all other variables in the model. These results indicate that the treatment had a direct effect on the two measures. Though negative and statistically significant, the coefficients appear to have a relatively small substantive direct effect.
If the treatment affects the gun-related indices by heightening the salience of racial attitudes and specifically racial resentment, we would expect the effect of the treatment to be contingent on the respondent’s score on symbolic racism.Footnote 11 Therefore, we expect that the effects of the prime on gun policy preferences and gun attitudes to be stronger among those respondents who score higher on symbolic racism than those who score lower on this measure of prejudice. As Table 1, columns 2 and 5 report, the effect of symbolic racism on the two indices is negative, statistically significant and substantively large. As reflected in Table 1 columns 3 and 6, both interactions of symbolic racism with the treatment are statistically significant at the p < 0.05 level (based upon a one-tailed test). And the signs of the relevant coefficients are such that both interactions indicate that the relationship between symbolic racism and each of the two indices is stronger (i.e., the coefficient is further from 0) for the treatment group.
These results provide supportive evidence of a causal relationship between racial considerations and gun policy preferences among whites. Our priming experiment shows that mere short exposure to pictures of blacks can drive opposition to gun control. Furthermore, the experiment suggests that these effects are stronger among those who express higher levels of racial resentment.
Cross-Sectional and Lagged Dependent Variable (LDV) Analyses of ANES Data
For the next set of analyses, we use several datasets: the 2004, 2008 and 2012 ANES studies and the 2013 ANES Recontact Study. All analyses include respondents who self-identify as non-Latino whites. Descriptive statistics for all datasets are presented in the Appendices.
Our key independent variable is the symbolic racism measure. The predictors for all analyses are identical. We draw our control variables from the relevant literature in political science, sociology and criminology. As noted earlier, this literature has focused on self-interest (variously conceptualized as gun ownership, fear of crime, or experience of victimization) as a major driver of gun policy preferences (Wolpert and Gimpel 1998; Hauser and Kleck 2013; Kleck et al. 2011). Gun ownership is our key predictor of self-interest. Because the ANES does not include measures of fear of crime, we include support for government spending for crime control as a proxy. Our assumption is that individuals who want higher spending on crime are more likely to fear crime than those who do not support higher spending on crime, ceteris paribus. Appendix B shows the items from which indices were constructed. Appendix C has descriptive statistics on all ANES variables included in our analyses.
Guided by extant literature on gun policy preferences, we control for a variety of values and beliefs such as egalitarianism, authoritarianism and a preference for small government. Political values such as beliefs in social hierarchy or more egalitarian attitudes have been implicated in driving the gun policy preferences of whites (Kahan and Braman 2003; Braman and Kahan 2006). Authoritarianism has been found to influence attitudes about criminal justice issues (Hetherington and Weiler 2009). Finally, studies find strong correlations between gun policy preferences, political ideology, and partisanship with conservatives and Republicans more likely to oppose gun restrictions (Wolpert and Gimpel 1998; Kleck 1996). We also control for key demographics including “South.” A number of studies have identified gun ownership as emblematic of a Southern “culture of violence” (e.g., Ellison 1991; Nisbett and Cohen 1996).
Starting in 2000, the ANES incorporated the following item: “Do you think the federal government should make it more difficult for people to buy a gun than it is now, make it easier for people to buy a gun, or keep these rules about the same as they are now?” Response rates among whites to this question for each ANES that included the item are reported in Appendix A.Footnote 12 We model responses to this item using an ordered logistic specification and we perform and report cross-sectional analysis for the three most recent studies (ANES did not include the “gun in household” item in the 2000 survey). Based upon our theory, we anticipate symbolic racism to be negatively associated with support for increased gun regulation.
Table 2, columns 1, 3 and 5 report the results of the ordered logit analyses based upon the three datasets that included the above specified dependent variable and the gun in household item.Footnote 13 Consistent with our expectations, the coefficient associated with symbolic racism is negative in each of the three analyses and statistically significant (p < 0.05 in 2008 and 2012; p < 0.10 in 2004).Footnote 14
The associations between gun regulation preferences and the control variables are largely consistent with past analyses. That the models’ results for the control variables are consistent with expectations and past findings reinforces the credibility of the estimated relationship between gun regulation preferences and symbolic racism.
In order to provide a sense of the relative substantive effect of each of the predictors on gun regulation attitudes, Table 2 reports the change in probability of a respondent selecting the “make it more difficult [to buy a gun]” response due to the maximum change in the associated predictor while holding all other predictors constant at their means.Footnote 15 For example, based upon the 2008 dataset, Table 2 column 4 indicates that going from the minimum to the maximum value of symbolic racism reduces the predicted probability of being in the top category by 0.25. This 0.25 decrease reflects the difference between the predicted probability of being in the top category when symbolic racism is set to its maximum value and all other variables are equal to their means minus the predicted probability of being in the top category with symbolic racism set to its minimum value and all other predictors set to their means (which are 0.32 and 0.57, respectively).
Columns 2, 4 and 6 of Table 2 indicate that egalitarianism and gun in household predictors had the greatest predicted substantive effect (with the mean of the maximum changes being 0.39 and −0.25, respectively). Symbolic racism, with estimated maximum effects ranging from −0.16 to −0.25 (and a mean of the maximum changes of −0.21) appears to be at the top of the second tier of predictors. The substantive effect of symbolic racism is slightly greater than the other predictors in the second tier such as gender, ideology and libertarianism (with the mean of the maximum change being +0.20, −0.20 and −0.19, respectively—though these differences in maximum change are not statistically significant). The magnitude of the effect associated with symbolic racism and its similarity with the magnitude of effect of other key predictors suggests that the relationship between symbolic racism and gun regulation preferences is substantively large.Footnote 16
The direction of the relationship between symbolic racism and gun regulation preference is very robust for all three analyses. Using weights or not and changing control variables do not affect the statistical significance of the relationship between symbolic racism and gun regulation preference in the 2008 and 2012 analyses. However, the statistical significance of the relationship between symbolic racism and gun regulation preference is not as robust based upon the analysis of the 2004 ANES dataset. The relative robustness of the inferences based upon the latter two datasets compared to the earlier two datasets could be due to idiosyncratic factors, such as sample size and mode of data collection differences (the 2004 ANES sample size was smaller than that of the latter two studies). Overall, our correlational multivariate analyses using the same dependent variable across the three studies suggest a strong negative relationship between symbolic racism and support for gun regulation.
The gun policy preference item modeled above was among the few items that were included in the 2012 wave and the 2013 recontact study. This allows us to estimate a model with an LDV on the right hand side (RHS) of the equation. Note that in this model all the predictor variables come from the 2012 wave and are identical to those used to estimate the models in Table 2. The only variable that has changed is the dependent variable: for this analysis we use the 2013 responses to the generic gun policy item. The 2012 responses to the item are used as the LDV on the RHS of the equation (Table 3). Including a LDV as a predictor helps minimize the probability that results are biased due to specification error since the LDV should “pick-up” the effects of any variables that influence attitudes towards the dependent variable but are not explicitly specified in the model (Achen 2000; Keele and Kelly 2006; Beck and Katz 2011). Though our primary rationale for including the LDV in the model is as a robustness check, including the LDV on the RHS of the model and using the policy preference at time t + 1 as the dependent variable reduces the likelihood that the dependent variable has an endogenous relationship with the predictors since opinion at time t + 1 is unlikely to influence opinion at an earlier moment in time.Footnote 17
The first two models presented in Table 3 report the results of an ordered logistic analysis and an OLS model without the LDV.Footnote 18 In general, the results from the ordered logistic analysis are consistent with the results of the full model based upon the 2012 ANES data reported in Table 1, column 3. The coefficients associated with symbolic racism are negative and statistically significant at the p < 0.05 level (one-tailed). Though the absolute value of the ordered logit coefficient is greater than the absolute value of this coefficient reported in Table 2, (−0.985 vs −0.653), the difference is not statistically significant at the p < 0.05 level.Footnote 19
The results of the OLS analysis without the LDV are provided as a baseline for the OLS analyses which include the LDV on the RHS of the equation. Note that the inferences based on the OLS results presented in Table 3, column 2 are almost indistinguishable from the inferences from the initial ordered logistic analysis as reported in Table 3, column 1. Furthermore, the signs of the coefficients remain negative regardless of model specification. This provides reassurance that the linear characterization is not overly distortive.
Table 3, model 3 reports the results of an OLS analysis including the LDV. Not surprisingly, the R2 increases substantively from 0.28 to 0.44.Footnote 20 A number of predictors that were statistically significant at conventional p-levels in the model without the LDV cease to be so in the LDV model. Of the coefficients that remain statistically significant at the p < 0.10 level (two-tailed), many decrease in magnitude. For our purposes, the key result is that the symbolic racism coefficient remains negative and significant (at p < 0.10 level, two-tailed).Footnote 21 Table 3, model 4 presents a more parsimonious version of the linear LDV model. Here, the symbolic racism coefficient is almost identical to that in the full model without the LDV (−0.260 vs −0.270) and statistically significant (at p < 0.05 level, two-tailed). This minimal decrease in the size of the coefficient with the inclusion of the LDV in the model suggests that symbolic racism is a robust predictor of gun regulation preferences.
We can assess the relative substantive effect of symbolic racism by comparing its maximum effect to that of other predictors.Footnote 22 Other than the LDV, no predictor has a greater maximum effect than symbolic racism. The maximum effects associated with all the other predictors are all closer to 0. In other words, in the final model with the LDV, the maximum effect associated with symbolic racism indicates that this variable is one of the most substantively important factors predicting attitudes toward gun restriction.
Discussion of Results of ANES Data
The multivariate analysis of the cross-sectional data demonstrates that symbolic racism is a statistically significant and substantively important predictor of gun regulation preferences pertaining to the regulation of firearms in the 21st century. By incorporating the lagged dependent variable into the model, the analysis of the 2012–2013 panel data indicates that these results are relatively robust to model specification (i.e., omitted variable bias) and helps address concerns regarding endogeneity. And, most importantly, the analyses of our priming experiment demonstrate that the relationship between racial prejudice and gun regulation preferences is recursive, not spurious, and that symbolic racism is the mechanism by which racial prejudice influences gun regulation attitudes.
Several limitations of the analyses deserve discussion. The analyses of the NES data use a single item to assess the public’s general attitude toward gun regulation. We doubt that a single item ever provides an optimal measure of the public’s attitudes on a given issue. The item’s reliability is suggested by the consistency/similarity of empirical results across the three datasets analyzed (i.e., the 2004, 2008 and 2012 NES studies). The statistical and substantive importance of the LDV as reported in Table 3 also suggests that the item is relatively reliable. The fact that the item correlates highly with additional gun policy items included in the 2012–2013 panel study and that an exploratory factor analysis indicates that all of these items load upon a single dimension provides some reassurance as to the item’s validity.Footnote 23 But future research regarding identity, prejudice and gun policy preferences would benefit from a more nuanced measure of gun regulation preferences (cf. our experimental study).
The issue of an underspecified model (i.e., omitted variable bias) can never be ignored. For example, we suspect that contextual factors such as the poverty rate, crime rate and various demographic characteristics of a respondent’s neighborhood might influence opinion. Consequently, our research agenda includes integrating such contextual factors into a multilevel model of gun regulation preferences. Fortunately, there is little reason to suspect that contextual factors will fundamentally change the individual level relationships identified above. Reassuringly, out LDV analysis suggests that the observed relationship between symbolic racism and gun regulation preferences will be robust to omitted variable bias. Furthermore, our priming experiment affirms that this relationship is a function of psychological processes at the individual level (though we suspect that the relevance and strength of such individual processes may be influenced by contextual factors).
Perhaps the most important limitation of the empirical analyses is their reliance upon data from the 21st century. Our theoretical expectation of a negative relationship between symbolic racism and gun regulation preferences rests upon an understanding of whites’ gun regulation preferences becoming caught-up in and intertwined with the narratives of the white resistance to the civil rights. The persuasiveness of this story would be greatly augmented by analyses of earlier data. Unfortunately, data from this earlier time period do not include measures of both gun regulation preferences and symbolic racism. However, the fact that the relationship between symbolic racism and gun regulation preferences remains so strong in the 21st century is a testament to the continued importance of racial prejudice and the influence of white resistance narratives on contemporary politics and policies in the United States.
Conclusion
This study attempts to make a number of substantive contributions to the literature. At its most basic, we demonstrate that racial prejudice influences white opinion regarding gun regulation in the contemporary United States. Our analysis suggests that scholars need to expand the set of issues considered racialized to include gun policy. Perhaps most importantly, we provide a causal story that explains the origins of the relationship between racial resentment and classical liberal narratives about rights and freedom, such as the gun rights discourse. This story emphasizes the plasticity of racialized boundaries and how historically contingent moments provide opportunities to (re)shape racial associations and meanings.
To be clear, the emergence of a rights narrative as part of a white lexicon of resistance to black (and even low class white) civil rights dates back to earlier eras (King 1999; Scalia 1998). However, the post-WWII era opened up the opportunity for this narrative to form the basis for the discourse of a number of illiberal white movements. During these earlier periods, we expect that there was widespread white support of gun regulations, especially restrictions of gun ownership targeting blacks. The legislative record up through the early 20th century strongly suggests that whites sought to keep firearms from the hands of blacks (Cottrol and Diamond 1995). We argue that the post-WWII era and the civil rights movement represent critical junctures when prejudiced whites started to shift away from supporting gun control as a result of the institutional and cultural changes of the era. Indeed, we strongly suspect that such a change in gun policy attitudes among whites was possible because guns have been a marker of white privilege throughout American history.
This theory has implications for how messages about gun control are communicated. For the past two decades the strategy of “gun control” advocates has been to focus on the rational public health benefits of additional gun regulations. In contrast, the messages propagated by “gun rights” advocates utilize the language and symbolism of rights along with the political and judicial tactics of the civil rights era to link gun ownership to traditional values and anti-black prejudice. Though the public debate on the issue of gun regulation reflects a two-sided information flow, the data indicate that over the past two decades public support for gun control has declined and has done so among whites in particular. A variety of factors may have contributed to this decline; however, it appears that the cost/benefit arguments and statistics used by “gun control” advocates have been no match for the emotional and persuasive power of gun rights messaging which invokes the white, gun-carrying every-person who defends home and Democracy against (nonwhite) bad guys.
We believe that racial prejudice colors all aspects of the debate regarding gun policy, including crime and its representations, and the role of government in society. Reshaping the gun regulation debate requires a deeper understanding of the relationships among racial prejudice, partisan politics, and the foundational but unconscious emotional role that guns play for a significant portion of the white population in America.
Notes
Kleck (1996) does not explain how his measure of racial attitudes was constructed, so it is difficult to evaluate his results.
We conducted analyses of the 2004–2012 ANES data using a difference in stereotypes measure and also a difference in affect (thermometers) measure. Neither measure was statistically significant in any of the models. This further suggests that racial attitudes linked to old fashioned racism are not driving whites’ opposition to gun control. Results are available upon request.
We excluded from the analysis the 26 individuals who completed the survey in less than 6 min or more than 90 min. The rationale for this exclusion was that ‘speeders” were not paying enough attention to the questions while those who took an excessive amount of time to complete the survey most likely did it in parts and these interruptions could have an effect on the data. Including these 26 individuals does not change the results.
In addition to gun policy questions, respondents were asked about other policy preferences, attitudes toward government, fear of crime, racial attitude questions, and a long demographics battery.
For example, the mean income of the subject pool was lower than that of the U.S. white population ($45,000 vs. $55,000, respectively).
We opted for using the IAT pictures as a racial prime, rather than administering the full IAT because of time limitations.
We were concerned that the gender of the images’ subjects might affect the treatment and its influence on subjects’ gun policy preferences. Consequently, half the treatment group received images of females only and the other half were exposed to images of males only. Interestingly, our analyses found no systematic differences between the effect of the two treatments. Consequently, for purposes of this paper, we treat exposure to either set of images as a single treatment.
Of the six policy proposals that did not fall on the first dimension, five fell on a second dimension (which had an eigenvalue of 1.15) and the sixth fell on a third dimension (which had an eigenvalue of 0.53). This third dimension item asked about immigrants. The five items that fell on the second dimension asked questions about safety rather than gun regulation per se. Examples of policy proposals that fell on this second dimension included: “More teachers and school officials with guns in schools,” “Teach school children how to use and operate guns safely,” and “Place armed guards in all public and government buildings and events (e.g., schools).”
Since the responses to the 15 policy and belief items were coded on a four point scale, we analyzed the effects of the treatment using ordered logit; but none of the inferences or substantive results based upon the ordered logit analyses differed from the inferences or substantive results based upon the OLS analyses. In order to simplify interpretation, we are reporting the OLS results. Ordered logit results are available upon request.
All dependent variables were recoded such that higher values indicate greater support for increased gun regulation. The key independent variable is binary (1 = treatment; 0 = control). We anticipate the coefficient associated with the treatment to be negative if the treatment primes racial attitudes and racial attitudes are negatively associated with attitudes towards firearms. In these analyses, we also control for age, gender, religion (Protestant), education, income, partisan identification and ideology. Including these control variables does not change the results. But we include the control variables to reassure readers who might be concerned about the characteristics of the subjects or about the randomization algorithm.
Our measure of symbolic racism is the standard used in the discipline (e.g., Sears and Henry 2003). See Appendix B for the items.
Based upon the ANES data, support for gun regulation among whites has declined during the 2000-2012 period. This is consistent with data from other surveys.
Though not reported, we performed an identical analysis of the 2000 ANES (sans the gun in household item). Inferences and substantive effects from this analysis were not meaningfully different from the results reported in Table 2. Results of an ordered logit analysis of the 2000 ANES are available upon request.
Consistent with standard practice in the field, included tables indicate statistical significance based upon two-tailed hypothesis tests. However, we think a one-tailed test is more appropriate given that our theory provides an anticipated direction of effect. For the analysis of the 2004 ANES dataset, symbolic racism is statistically significant at the p < 0.05 level based upon a one-tailed test.
Technically, we could provide such a change in probability table for each response category. However, providing such additional information seems to us overly burdensome for the reader. Moreover, since so few cases (4–6 %) fall into the “make it easier” response category, the probabilities of being in the bottom category are very small for the overwhelming percentage of cases, so changes in the probability of being in this category due to a maximum change in any one predictor is quite small. In other words, most of the “action” is in changes in the probability associated with being in the “keep these rules about the same” and the “make it more difficult … to buy a gun” categories. Also, since the changes in probability associated with being in the top category tend to be greater than the changes in probabilities associated with a respondent selecting either of the other two response categories, we think that providing the change in probability associated with being in the top category provides an analysis most akin to the standard understanding of maximum effects.
Based upon a Wald test, we find that the analyses of the 2004 and 2008 ANES datasets are consistent with the parallel regression assumption, but that this assumption is not met by the ordered logit analysis of the 2012 ANES dataset. However, the relationship between symbolic racism and gun regulation preferences as reflected in the 2012 ANES dataset appears robust to model choice. For example, a multinomial probit analysis of the 2012 ANES dataset using the identical variables yields two coefficients associated with symbolic racism, both of which are in the anticipated direction and both of which are statistically significant at the p < 0.05 level (two-tailed). We discuss the results of multinomial probit analyses rather than multinomial logit analyses because the former is not dependent on the independence of irrelevant alternatives (IIA) assumption. Multinomial results are available upon request.
Analysts have noted a few disadvantages to including an LDV on the RHS. For example, in the context of time-series analysis, Achen (2000) notes that including a LDV on the RHS often biases estimates downwards. However, we are primarily interested in checking the robustness of our inference regarding the relationship between racial resentment and gun policy preference. Also, we find that the includsion of the LDV on the RHS does not reduce the estimated coefficient associated with symbolic racism by more than 17 % (see Table 3).
Issues regarding the consistency of maximum likelihood estimates of logistic regression using panel data have long been recognized (Hsiao 1986). Even when possible, consistent estimates of fixed effects logistic models involve the loss of significant information since only those cases in which the dependent variable changes state contribute to the log-likelihood function (and note that a fixed effects model is not possible given that T = 1 for the predictors). Because of these concerns, development economists often utilize the LPM as the least worst option (for a discussion, see: Beck 2011). Given these issues, we have opted to do the same.
Though the difference in coefficients is not statistically significant, it is possible that racial attitudes had a stronger relationship with gun policy preferences in July 2013 than in November 2012. In the middle of January 2013, less than a month after the Newtown, CT massacre, Obama announced 23 executive actions his administration would take in regards to gun control and he asked Congress to pass several gun control measures. Congress debated a variety of measures and the Senate voted on a series of gun control measures in late April 2013. It is conceivable that the push by Obama’s administration and Congressional Democrats on gun control during this period might have made racial attitudes more salient on this issue among the public.
Indeed, a simple bivariate AR(1) model accounts for approximately 38 % of the variance of the dependent variable.
Technically speaking, the coefficient associated with the lagged dependent variable can be interpreted as would any other coefficient using an OLS specification. A different approach would be to think of the model as a panel analogue to a “dead start” time series model. In this model, the coefficient associated with a lagged dependent variable can be thought of as an estimate of a decay rate that can be used to help parse the short-term and long-term effects of a change in the predictor on the dependent variable. In this approach, the −0.260 can be thought of as the short term change in support for “making it more difficult to purchase a gun” due to a one unit increase in symbolic racism. But the long term effect of such a one unit increase in symbolic racism would be −0.576 (which is simply \((\beta_{0} + \beta_{ 1} )/\left( { 1{-}\delta } \right) = (0 \pm 0. 2 60)/\left( { 1{-}0. 5 4 9} \right)\)).
Maximum effect of \({\text{X}}_{k} = {\text{b}}_{k} \times ({ \hbox{max} }.{\text{ value of X}}_{k } - { \hbox{min} }.{\text{ value of X}}_{k} )\).
Factor analysis are available upon request.
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Acknowledgments
The authors would like to thank Kate Boulay, Michelle Boyd, Jennifer Carlson, Steven Corey, Amanda D’Urso, Deborah Holdstein, Richard Johnson, Lisa Miller, Dionyssis Mintzopoulos, Michael L. Owens, Davin Phoenix, Beth Richie, Linda Skitka, Vesla Weaver, Dick Winters, Cara Wong, and the four anonymous reviewers for valuable comments and suggestions. This work was supported with grants from the Political Science Department, the Office of Social Science Research (OSSR) and the Chancellor’s Discovery Grant program at the University of Illinois at Chicago. Dr. Dick Simpson also provided generous financial support from his grant and we would like to thank him for that. Dr. Filindra benefited greatly from a fellowship by the Institute for Research on Race and Public Policy also at UIC. We also want to thank the American Political Science Association’s Public Policy Section who awarded a version of this article with the best paper award for 2014.
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Filindra, A., Kaplan, N.J. Racial Resentment and Whites’ Gun Policy Preferences in Contemporary America. Polit Behav 38, 255–275 (2016). https://doi.org/10.1007/s11109-015-9326-4
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DOI: https://doi.org/10.1007/s11109-015-9326-4