Abstract
Since the advent of public opinion polling, scholars have extensively documented the relationship between survey response and interviewer characteristics, including race, ethnicity, and gender. This paper extends this literature to the domain of language-of-interview, with a focus on Latino political opinion. We ascertain whether, and to what degree, Latinos’ reported political attitudes vary by the language they interview in. Using several political surveys, including the 1989–1990 Latino National Political Survey and the 2006 Latino National Survey, we unearth two key patterns. First, language-of-interview produces substantively important differences of opinion between English and Spanish interviewees. This pattern is not isolated to attitudes that directly or indirectly involve Latinos (e.g., immigration policy, language policy). Indeed, it emerges even in the reporting of political facts. Second, the association between Latino opinion and language-of-interview persists even after statistically controlling for, among other things, individual differences in education, national origin, citizenship status, and generational status. Together, these results suggest that a fuller understanding of the contours of Latino public opinion can benefit by acknowledging the influence of language-of-interview.
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Notes
For an overview of English/Spanish translations of political surveys, see Pérez (2009). We later return to this issue when we examine the links between item wording, measurement error, and opinion gaps by language-of-interview.
It is plausible the consequences of language-of-interview choice are similar to those that sometimes emerge when there are differences between survey respondents and non-respondents. In the latter case, individual differences in response/non-response can bias the results of opinion polls (Groves 2006). Yet research using survey-experimental designs finds that individual differences in response/non-response produce negligible amounts of bias in survey results (e.g., Curtin et al. 2000; Keeter et al. 2000, 2006). Determining whether something similar occurs in the realm of language-of-interview choice must await the development of comparable research designs.
Other bilingual surveys of Latinos (e.g., 1999 Washington Post survey of Latinos) find that closer to 40 % of respondents choose a Spanish interview. de la Garza et al. (1992) suggest their higher proportion of Spanish interviews may be due to an oversample of Puerto Ricans and Cubans in the LNPS. Other possibilities that de la Garza et al. do not discuss are the LNPS’s high response rate by today’s standards and its face-to-face mode, which may yield a higher proportion of Spanish interviews. A third possibility is that the underlying language use patterns of Latinos may have changed over time (i.e., more English-speaking Latinos in 1999 than in 1989).
Interviewer characteristics are unavailable in the LNPS, but de la Garza et al. (1998) note that 159 interviewers were trained, 138 interviewers conducted one or more interviews, and five-sixths of this pool was bilingual.
The actual sample used is N = 2,616 (645 Cubans, 1,428 Mexicans, and 543 Puerto Ricans).
LNPS Spanish interviewees are older, less educated and wealthy, less likely to own a home and to be employed, and more likely to be married. More than 90 % and less than 50 % of Spanish interviewees are foreign-born and citizens, respectively (Table B, online appendix). Per Sullivan et al. (1979), differences in age, education, income, and foreign-born and citizenship status are likely correlates of one’s ability to articulate political opinions, as well as language-of-interview. Our subsequent analyses include these individual differences as covariates. For more insight on Latino language pattern variation, see Lopez (1996), Portes and Schauffler (1996), and Stevens (1985, 1992).
The LNPS codes for six racial categories that do not map onto most survey race categories: White, black, other, Spanish label (e.g., Hispanic, Mestizo), Color oriented (e.g., Moron, Triune, Brown), and Race label (e.g., mulatto, North American). In Tables 2, 3, 4, we simply create a dummy variable for Whites (1 = Whites, 0 = all others).
Moreover, as we report later in the paper, the conclusions we draw from our analysis of these three items remain unchanged if we broaden our focus to other items in the LNPS, as well as items in other datasets (e.g., the LNS).
These values were calculated via Clarify (King et al. 2000), while holding other covariates at their means.
The unequal chances item (Table 1) combines respondents who “somewhat” and “strongly” disagree that unequal chances in life are all right.
The values in Table 5 reflect the magnitude of interview language differences, while additively controlling for each plausible account (e.g., demographic characteristics, language use and proficiency).
Results are based on a pooled ANES sample of Latinos from 1978 to 1998 (N = 1,438). LNPS respondents give fewer non-response answers, but this might arise from differences in survey mode (face-to-face vs. telephone interviews). When ANES items are compared to like items in a Latino telephone survey (e.g., the 1999 Washington Post poll or the 2002 Pew Hispanic Center poll), no differences in non-response emerge.
216 is the upper limit of cases on the items we examine, since a non-trivial number of observations are excluded due to ambivalent or non-compliant responses. There are several limitations with this kind of analysis, of course. One is that our measure of language familiarity could have significant measurement error, if substantial portions of our respondents coded “Spanish primary” or “English primary” in fact chose to be interviewed in a language other than the one they are most familiar with (23 % of Spanish primary respondents chose to interview in English, 47 % of English primary respondents chose to interview in Spanish). Second, language proficiency and usage may reflect distinct constructs. If we focus just on language proficiency, reliable gaps by interview language remain in 35 out of 44 items for Spanish primary respondents, 43 items for bilingual respondents, and only 5 items for English primary respondents (N = 483, 2,056, and 107, respectively).
When translating questionnaires, scholars strive to ensure similar meaning across languages (i.e., functional equivalence) (Pérez 2009). This often entails some sacrifice in strictly identical wording between items, since a one-to-one correspondence might fail to achieve uniform meaning (e.g., two languages may not have the same analog for a common concept). Critically, the LNPS and LNS appear to generally display functional equivalence. First, the second author in this paper is a native English and Spanish speaker. His examination of LNPS and LNS items detected a few idiosyncratic differences, which we note above in the text. Minor discrepancies like these are common when aiming for functional equivalence. Second, our conclusion about the uniform meaning of these translated items was further corroborated by an independent professional translator, who also failed to identify egregious translation differences that could undermine the goal of functional equivalence.
de la Garza et al. (1998) used a focus group approach to translate the English/Spanish items in the LNPS. Here the English instrument was designed by the Principal Investigators (PIs). The Spanish version was drafted by a Chilean Spanish speaker. The English questionnaire was then vetted against its Spanish analog by a focus group of six bilingual and college-educated individuals (2 Mexicans, 2 Puerto Ricans, 2 Cubans). Our analysis also uses the 2006 Latino National Survey (LNS) led by Fraga et al. Here the PIs prepared the survey in English and then translated it themselves into Spanish (several of the PIs are native or fluent Spanish speakers). This bilingual instrument was then evaluated by the study’s advisory board and graduate students, both of whom were native or fluent Spanish speakers. The PIs resolved any remaining discrepancies. The final instrument was field tested via a small pilot study. No further anomalies emerged (personal communication with LNS PI Michael Jones-Correa, October 24, 2012; personal communication with LNS PI John Garcia, October 29, 2012).
We can also rule out that language-of-interview gaps reflect other individual differences in respondent characteristics, rather than the interactive nature of the survey interview. This version of the social desirability thesis reduces to a variant of the “omitted variable bias” thesis, but we have already seen that language-of-interview gaps remain even after controlling for a large number of individual differences among Latino respondents.
The possibility of language-of-interview differences as a window into assimilation patterns and ethnic identity formation is a topic pursued more explicitly by Lee (2000).
In the interest of space, we have centered on these three items. Yet the link between interview language and Latino opinion broadly extends to other LNPS items. To show this, we used a random number generator to randomly select and report an additional eight (8) items, or 20 % of the remaining forty-one (41) in Table 1. A reliable association between language-of-interview and Latino opinion also emerges in six (6) of these eight (8) items, with five (5) of them achieving significance at the 5 % level and one (1) at the 10 % level. The coefficients/standard errors for these items are: (1) Love for U.S. (.23/.07); 2) Spending on science (−.21/.08); 3) Spending on the environment (.09/.09); 4) Cubans: Feel Therm. (−3.08/1.66); 5) Mexican Americans: Perc’vd Discrimination (.04/.07); 6) Asians-Perc’vd Discrimination (−.20/.07); 7) Teach U.S. history (.31/.08); and 8) Central American unrest (−.16/.08). All estimates are from ordered probit regressions, except those for item 4, which were yielded via OLS.
Intra-language heterogeneity is less of a problem when polling Latinos in subnational contexts where specific national origin group(s) predominate (e.g., Mexicans in Texas). Yet even in these cases, it still behooves scholars to pay greater attention to the resemblance between item translators and the population being sampled.
The mean interview was 40.6 minutes. The cooperation and response rates for this survey were 36.7 and 9.54 %, respectively (personal communication with LNS co-PI Gary Segura, March 2010).
The smallest national origin group in our analysis is Dominicans (n = 335). This group of cases still permits a meaningful analysis. The next smallest group excluded from our analysis is Guatemalans (n = 149). Here, if we were to include this group and find a reliable association, it would be hard to say whether it stems from being Guatemalan or something unique about these Guatemalans. Thus, we focus on the five groups described above.
The wording for these items can be found in Table A (online appendix).
In the interest of space, we define each construct here and refer readers to relevant scholarship. These citations are illustrative, but by no means exhaustive. Political knowledge refers to individuals’ factual information about politics (Delli Carpini and Keeter 1996). Greater political knowledge, among other things, increases people’s engagement with politics. Latino attachment refers to identification with other Latinos. Formally, it refers to one’s identification with the social object, Latino (Turner et al. 1987). A key component of group attachment is a sense of commonality with it (Garcia 2003; Garcia Bedolla 2005). It is theorized the measures at hand capture this sense of attachment. Americanism refers to Latinos’ sense that national identity is ethno-racially delimited (e.g., Higham 1981; King 2000). Put differently, this construct captures the degree to which an individual sees the boundaries of national identity as ethno-racially impermeable. Thus, the more intense these perceptions, the less relevant American identity should be for one’s political behavior. On American identity and Latinos, see Fraga et al. (2010) and Citrin et al. (2007). On identity permeability, see Jackson et al. (1996) and Lalonde and Silverman (1994). Finally, perceived intergroup competition refers to one’s sense of conflict, across several domains, with an outgroup relative to one’s ingroup. Recently, this construct has been used to study Latino perceptions of political and economic competition with Blacks (Barreto et al. 2010, 2011; McClain et al. 2006, 2011). This construct is generally measured by subtracting one’s perceptions of competition with Latinos from one’s perceptions of competition with Blacks. We examine the items attending each set of perceptions.
See tables D through G (online appendix) for the relevant confirmatory factor analyses (CFAs).
In its raw metric, this item ranges from 1 to 5, with higher values indicating lighter phenotype.
Previous research finds this variable can, in some cases, affect Latinos’ survey response (Garcia 2010).
In fact, if we examine comparable items from the LNS and LNPS, this conclusion remains unchanged. While these two surveys do not share identical items, we did identify eight (8) LNS items that approximate those in the LNPS. Our tests uncover a reliable association between language-of-interview and Latino opinion for seven (7) out of eight (8) of these LNS items. See table H (online appendix) for the relevant items and statistical results.
Results are available from the authors upon request. In most instances, language-of-interview appears to be a more decisive influence on Latino mass opinion than does ethnicity-of-interviewer.
See Lee (2001).
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Acknowledgments
Author order is alphabetical. The authors thank Mike Alvarez, Darren Davis, Zoltan Hajnal, Cindy Kam, Paula McClain, Natalie Masuoka, Karthick Ramakrishnan, Ricardo Ramírez, and Lynn Sanders for valuable discussion and comments on this project. The authors are also grateful to the Editors and three anonymous referees for constructive advice on this article.
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Lee, T., Pérez, E.O. The Persistent Connection Between Language-of-Interview and Latino Political Opinion. Polit Behav 36, 401–425 (2014). https://doi.org/10.1007/s11109-013-9229-1
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DOI: https://doi.org/10.1007/s11109-013-9229-1