Consistent with theories of modern racism, we argue that white, non-Hispanic Americans have adopted a “coded,” race-neutral means of expressing prejudice toward Hispanic immigrants by citing specific behaviors that are deemed inappropriate—either because they are illegal or threatening in an economic or cultural manner. We present data from a series of nationally representative, survey-embedded experiments to tease out the distinct role that anti-Hispanic prejudice plays in shaping public opinion on immigration. Our results show that white Americans take significantly greater offense to transgressions such as being in the country illegally, “working under the table,” and rejecting symbols of American identity, when the perpetrating immigrant is Hispanic rather than White (or unspecified). In addition, we demonstrate that these ethnicity-based group differences in public reactions shape support for restrictive immigration policies. The findings from this article belie the claim of non-prejudice and race-neutrality avowed by many opponents of immigration.
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Allport’s (1954) original definition of prejudice also included the condition that the antipathy be “based upon a faulty and inflexible generalization.” However, as Brown (2010) and Stangor (2009) note, most scholars have subsequently dropped that requirement from the operational definition of prejudice.
Excerpts accessed from Numbers USA’s organizational website: https://www.numbersusa.com/content/. Emphasis in italics were added by authors and do not appear in the original quote.
Quote from Alabama State Representative Kerry Rich. Retrieved from: http://www.sandmountainreporter.com/news/local/article_b58d77c2-9920-11e0-8a09-001cc4c03286.html.
Quote from Alabama State Senator Scott Beason. Retrieved from: http://www.wncftv.com/localnews/Bentley-Signs-Illegal-Immigration-Reform-Bill-into-Law-123587664.html.
Excerpt taken from an interview given by Rubio on Democracy Now! on October 20, 2011. Retrieved from: http://www.democracynow.org/shows/2011/10/20.
In our usage, the terms “Whites” and “white Americans” exclude individuals who would identify themselves as ethnically Hispanic.
Data collected by the Survey Research Center, University of California, Berkeley, from February 1 to November 21, 1991.
A list-assisted method of random digit-dialing (RDD) was used to obtain phone numbers in the sample from all 48 contiguous states, including the District of Columbia. Within selected households, individuals 18 years and over were chosen at random for participation. Multiple attempts were made at each contact number (as many as seven attempts) in order increase response rates and give potentially eligible respondents a reasonable opportunity to participate in the survey. Moreover, households and individuals who were initially unwilling to participate in the survey were contacted multiple times in an attempt to persuade them to participate. Calls were staggered over times of day and days of the week to maximize the chance of making contact with potential respondents. In total, 6,032 telephone numbers were dialed, each of which was given a final disposition: 3,763 numbers were deemed ineligible (e.g., nonworking, businesses, etc.), 1,109 numbers were of unknown eligibility (always busy, never answered, etc.), and the remaining 1,160 numbers were coded as eligible households (275 completes, 304 refusals, 36 non-whites, 70 language unable, and 475 callbacks). We used two methods of determining levels of participation in this survey: (1) The Cooperation Rate (AAPOR Formula #4) was 51.4 %; and (2) the Response Rate (AAPOR Formula #4) was 22.6 %. The response rate is a very conservative estimate of participation, while the cooperation rate adjusts for the fact that many phone numbers in the list are non-eligible.
The “Pew Research Center Poll: Immigration” was sponsored by the Pew Research Center for the People & the Press and the Pew Hispanic Center. A total of 6,003 surveys were completed between February 8th and March 7th, 2006. The 2008–2009 American National Election Panel Study consists of an Internet panel of 4,240 Americans recruited via RDD sampling methods.
Percentages do not total 100 % because of rounding.
For ease of interpretation, we recoded this and all subsequent variables from 0 to 1.
For instance, a respondent who was presented with the scenario of a Mexican immigrant overstaying his or her visa was later asked to evaluate the seriousness of this same Hispanic immigrant working without paying income taxes.
Unlike the previous realistic experiments, our symbolic experiments involved only two conditions, and respondents were randomly assigned to either condition for each individual symbolic experiment. Our decision to exclude the race-neutral control condition for the symbolic experiments stemmed from our concern that the prior exposure to group cues in the realistic experiments could prime respondents in the neutral condition. Thus, participants in an unidentified symbolic condition, had we included one, could conceivably be primed to think of that “someone” as Mexican, British, or truly undefined, which would undermine the integrity of contrasts in responses to respondents assigned to the explicit Mexico or Canada conditions. To eliminate this possibility, our symbolic experiments explicitly specify the ethnic origins of the immigrant in question. In addition, we changed the ethnicity of the non-Hispanic immigrant in our symbolic experiments from British to Canadian to demonstrate that our results hold across different white immigrants.
As there were no significant differences in main effects between the non-Hispanic treatment conditions in our realistic experiments (i.e., British and undefined cues), we opted to combine them into a single category. Thus, we used a dichotomous group cue variable for our analyses (1 = Mexico; 0 = non-Hispanic). The coding for the symbolic experiments was similar (1 = Mexico; 0 = Canadian).
Perceptions about the illegal immigrant population of Hispanic origin came from responses to the following question: “If you had to guess, what percentage of the Hispanic immigrant population is living in the U.S. without legal documentation?” Responses ranged from “0” to “100 %,” with a mean of 42.3 % and a standard deviation of 24.6 %. Education is a 6-point scale, where a graduate degree serves as the highest category. Household income is a 8-point scale based upon $20,000 increments, and missing values were imputed in Stata based upon gender, education, age (and its squared term), and employment status. For ease of interpretation, all variables were recoded from 0 to 1.
Kinder and Kam (2010) demonstrate that ethnocentrism, or “a predisposition to divide the human world into in-groups and out-groups” (p. 8), strongly predicts anti-immigrant sentiment. They argue that individuals are predisposed to favor their ingroup at the expense of outgroups, and that antipathy toward outgroups should increase as a function of the cultural, linguistic, and ethnic distance of an outgroup to one's ingroup. According to this approach, the operative mechanism underlying our experimental findings could be general aversion to outgroups and “prejudice broadly defined” (Kinder and Kam 2010, p.52), rather than group-specific prejudice toward Hispanics divorced from an encapsulating ethnocentrism. While this alternative and more general framework could account for findings such as ours, this hypothesis is directly challenged by evidence that specific attitudes toward Hispanics, not ethnocentrism, influence immigration policy preferences (Valentino et al. 2013). In light of these countervailing findings, we should note that the primary goal of this article is to test for the existence of bias toward Hispanics by determining whether individuals evaluate the transgressive behaviors of Hispanic immigrants more negatively than those of non-Hispanic immigrants. Adjudicating whether the demonstrated bias in our experiments stems from prejudice toward Hispanics embedded within general ethnocentrism is beyond the scope of this article.
Once again, we opted to collapse the British and undefined treatments from our realistic experiments because we found no significant differences in main effects between these conditions.
To estimate the mediated effects of our group cue treatments on policy preferences, we used the mediation package in R (Tingley et al. n.d.) to regress (1) a continuous measure of perceived offensiveness of a given violation on a dichotomous group cue treatment variable using OLS, and (2) a categorical immigration policy item on the perceived offensiveness of a given violation, as well as a dichotomous treatment variable, using probit or ordered probit link functions.
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We would like to thank Dan Hopkins, Shanna Pearson-Merkowitz, Nick Valentino, and Cara Wong for their helpful suggestions on earlier versions of this article. We also thank Dustin Landers for his help during the data collection phase of this project.
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Hartman, T.K., Newman, B.J. & Scott Bell, C. Decoding Prejudice Toward Hispanics: Group Cues and Public Reactions to Threatening Immigrant Behavior. Polit Behav 36, 143–163 (2014). https://doi.org/10.1007/s11109-013-9231-7