Perceiving Immigrants as a Threat: A Motivational Approach to False Consensus

Abstract

False consensus, or biased projection of one’s opinion onto others, has repeatedly been described by political communication scholars as a derivative of selective exposure to attitude-consistent information. This study proposes a distinctive approach to understanding the phenomenon by suggesting “perceived threat” as a motivational factor that contributes to self-serving estimates of public opinion. Based on a nationally representative sample of U.S. adults, we test a path model in which political ideology relates to false consensus regarding the issue of immigration through cognitive assessments of communication environment and perceived immigration threat. Results suggest that the relationship between cognition and false consensus may not be direct but instead works through motivational factors when one perceives threat, and that conservatives are more sensitive to outgroup threat and thus are more likely to overestimate public consensus for their attitudes on immigration than their ideological counterparts. Implications of these findings are discussed.

Keywords

false consensus perceived threat immigration public opinion perceived media bias

The topic of false consensus holds an exclusive place in the field of political communication as it highlights the capacity of news media environment and interpersonal communications to influence citizens’ perception of social norms and public opinion climates. Widely conceptualized as the tendency to overestimate the commonness of one’s own views (Krueger & Clement, 1994; Ross et al., 1977), false consensus has been extensively cited as one of the outcomes of selective exposure (Marks & Miller, 1987; Tsfati et al., 2014) and ideologically likeminded communication networks (Wojcieszak, 2010). The recurring argument is that people tend to be attracted to information sources that are consistent with their political views, and selectively exposing oneself to consonant opinions may shield her/him from differing opinions, which subsequently results in egocentric biases in social perceptions. Previous studies suggest that ideologically homogeneous online forums may exacerbate false consensus (Wojcieszak, 2008), and that encountering disagreement could mitigate the extent to which people overestimate public support for their own views (Mutz, 2002).

While such studies provide useful insights into the relationship between interpersonal/mass-mediated communications and public opinion perceptions, they largely explain the false consensus phenomenon from the cognitive availability perspective—that is, how individuals employ the availability heuristic from an ideologically biased sample of information to estimate the prevalence of their own views. Considering the complex communication practices of citizens that expose them to diverse, or even opposing, political perspectives (Dubois & Blank, 2018), the cognitive explanation may not wholly address the motivational underpinnings to the consensus bias. Social scientists indicate “self-serving motive” as a key determinant of false consensus (Sherman et al., 1984) and argue that one’s motivation to seek support for her/his own position, along with cognitive availability, may jointly explain overestimation of consensus (Krueger & Clement, 1997). “Perceived threat” serves as one of the most prevalent motivational determinants for the false consensus effect; people may often unknowingly exaggerate the perception of agreement with others when they perceive threat to the self, or to their group, to protect the ego and justify their views (Sherman et al., 1984). From this viewpoint, overestimation of consensus works as a “self-defense or self-enhancement mechanism” to minimize perceived threat (Mannarini et al., 2015). We believe that the discussion of motivational factors in false consensus is particularly relevant in the political realm, given that people are “highly motivated to perceive the world in ways that satisfy their [political] values” (Jost et al., 2003, p. 341).

The principal goal of the current study is to examine the role of motivational factors, notably perceived threat that stems from cognition, in fostering false consensus effects. Although some have explored the motivational differences between conservatives and liberals to explain the association between ideology and false consensus (Stern, West, & Schmitt, 2014), we aim to advance the existing literature by proposing and testing a mediation model that identifies both cognitive (perceived opinion climate) and motivational factors (perceived threat) through which political ideology relates to self-serving estimates of public opinion. To explore the link between perception of threat and false consensus, we conducted a national survey of U.S. adults, with specific aims to measure citizens’ opinions regarding “immigration” and their estimates of the percentage of Americans who agree with their opinions on immigration. The issue of immigration has repeatedly been studied by communication research on perceived threat (e.g., Watson & Riffe, 2013), but still very little is known about how perceiving immigrants as a threat can lead one to overestimate public support for her/his opinions on immigration.

Cognitive and Motivational Explanations for False Consensus

False consensus is described as a special form of social projection, which refers to the tendency to expect others to be similar to oneself (Bauman & Geher, 2002; Robbins & Krueger, 2005). The cognitive explanation to consensus estimation (Mullen & Hu, 1988) posits that people rely on information that is more readily accessible from their memory when estimating the prevalence of attitudes/opinions in the population. Such information tends to be examples of agreement rather than disagreement, since individuals have the tendency to connect with others who are similar to themselves (Mannarini et al., 2015; Wojcieszak, 2008). From a mass communication perspective, selectively exposing oneself to likeminded media content may result in the impression that public opinion is more supportive toward her/his own point of view than it actually is (Wojcieszak, 2011). The cognitive availability approach, therefore, describes social projection as a judgmental heuristic that allows one to make quick decisions about social reality using news media and those whom they talk with as sources of information (Bauman & Geher, 2002). And because the estimates of consensus are often based on a biased and restricted sample of the broader population, they are prone to inaccuracy and therefore are erroneous and “false” (Mullen & Hu, 1988). Although erroneous estimation of public opinion can work in both directions, the term false consensus has generally been adopted to describe a phenomenon in which people wrongly assume that their attitudes or beliefs are shared by others (e.g., Krueger & Clement, 1994). Accordingly, this study conceptualizes false consensus as an overestimation of public support for own views.

The cognitive approach can be useful in explaining how personalized digital communication environment could contribute to false consensus by promoting exposure to likeminded views (e.g., Wojcieszak, 2008). Such explanations, however, often rely on findings obtained from a small number of online platforms, which may not fully encapsulate the variety of potential mechanisms underlying false consensus phenomena in today’s complex multimedia environment. The increasingly diverse communication repertoires of individuals—ranging from the use of multiple traditional/online media sources to discussing politics with friends and acquaintances—can actually expose people to a variety of political views, including those that do not align with their own political perspectives (Dubois & Blank, 2018). Recent studies on perceived media bias and hostility in public opinion (e.g., Ardèvol-Abreu & Gil de Zúñiga, 2017) suggest that citizens are not incapable of knowing the prevalence of conflicting political viewpoints, and that the use of ideologically homogeneous information sources would not completely trap people inside echo chambers that blind them from opposing views (Flaxman et al., 2016). The fact that overestimation of public consensus is an empirically robust phenomenon (Choi & Cha, 2019; Gershoff et al., 2008) begs the question of what factors might mediate the effect of “perceived political disagreement” on false consensus.

In that case, social scientists argue that one’s cognition (i.e., subjective perception of public opinion) may not directly lead to inflated estimates of consensus, but may rather work through various self-serving motives such as willingness to seek social support (Marks & Miller, 1987). From a motivational perspective, social projection serves as a cognitive means to fulfill one’s needs to obtain support and validation for attitudes, opinions, or behaviors. Perceiving that one’s own position is shared by others increases certainty regarding the correctness of this position (Schulz et al., 2020, p. 207), and by overestimating consensus, one might be “reassured of the normality and appropriateness” of her/his position (Hoorens, 1993). While the cognitive approach to social projection posits that consensus estimation results from “statistical” reasoning that considers all available sample information to make estimates, the motivational explanations stress the nonstatistical reasoning underlying one’s projection, highlighting “a sense of special value of self” (Krueger & Clement, 1994, p. 607).

This does not mean, however, that the cognitive and motivational explanations conflict with one another. The false consensus literature suggests that both availability heuristic and self-defense/enhancement motives are involved in erroneous consensus estimation, though one type of social projection mechanism might predominate over the other depending on people’s actual or perceived group status (majority or minority) (Krueger & Clement, 1997). Self-focused motives, in particular, are believed to exert greater influence over consensus overestimation than cognitive/informative factors when individuals perceive themselves as minorities, regardless of their actual status. Earlier findings show that those perceiving minority positions within a group or society (e.g., preferring unpopular activities over the popular ones) are more likely to overestimate the commonness of their positions than people perceiving majority status (Marks & Miller, 1987; Mullen & Hu, 1988). Egocentric estimation of consensus may thus satisfy the need to seek validation and support from others, as well as “the desire to escape the perception of being in the minority” (Sanders & Mullen, 1983, p. 59; Sherman et al., 1984). With that said, the hypothesized effect of perceived group status on biased consensus estimation (Krueger & Clement, 1997) suggests that cognitive and motivational reasoning processes may occur sequentially, rather than simultaneously, in a linear fashion—that is, one’s perception of her/his majority-minority status (or agreeable-hostile state of public opinion toward one’s views) leading to relevant self-serving motives that ultimately result in false consensus.

The notion that individuals are intrinsically motivated to minimize “perceived threat” (VanDellen et al., 2011) is particularly applicable to the discussion of false consensus stemming from a potential causal flow from cognition to self-centered motivation. Threat in false consensus literature is often conceptualized as an experience that evokes “discrepancies between desires or expectations and reality” (VanDellen et al., 2011, p. 52), referring to perceived rather than actual threat. Scholars argue that one may sense a threat to self when her/his own cognitive assessment of the situation calls for needs to seek social support, security or legitimacy (Marks & Miller, 1987; Schulz et al., 2020), and experimental findings suggest that such needs could lead to errors in social judgment (Sherman et al., 1984, p. 136). Morrison and Matthes (2011) found that social exclusion cues could place people under threat by triggering their need for social support, which in turn could cause them to overestimate consensus for their opinion. In this sense, exaggerated consensus for personal views/opinions serves as a self-defense mechanism—a reaction to perceived threat that results from individuals’ perceptions of social reality.

From Political Ideology to Perceived Threat—Through Perceptions of Social Disconfirmation

The perceived target of threat could either be to the self, or the ingroup one identifies with (Krueger & Clement, 1997; Sherman et al., 1984), and previous works have shown that perceived threat in a political context often reflects concern for an ideological ingroup rather than an individual’s ego. Those who identify with a political party tend to believe that their party’s values and norms are objectively true (Wirtz et al., 2016, p. 76), and they may feel threatened when an ideologically dissimilar outgroup challenges their group’s identity, beliefs and attitudes—that is, when they perceive group differences in what they believe (or wish to believe) to be universally right (Rios et al., 2018; Wirtz et al., 2016).

Political scientists provide ample evidence that political ideology could act as a significant predictor of the level of perceived threat discussed in false consensus literature. More explicitly, they argue that political conservatives are often more likely than liberals to perceive the outgroups as a threat to ingroup’s beliefs (Janoff-Bulman, 2009; Jost et al., 2003). Researchers note “opposition to change” to be one of the distinguishing motivational aspects of conservative ideology, which contrasts liberals’ accepting attitudes toward social change (Jost & Amodio, 2012; Proulx & Brandt, 2017). Wilson (1973) defines political conservatism as “the tendency to prefer safe, traditional and conventional forms of institutions and behavior,” and research findings in physiology, experimental psychology and neuroscience largely side with such definition—Oxley et al. (2008), for instance, showed that conservatives tend to be more physiologically reactive to threatening stimuli, while Kanai et al. (2011) found that the volume of the brain area responsible for threat-detection (right amygdala) is positively associated with political conservatism.

The idea that conservatives tend to be more sensitive to external threats than liberals, and thus are more likely than liberals to support socially protective policies that reduce threats (Oxley et al., 2008), is perhaps best reflected in research on public attitudes toward immigration. Immigration has been “one of the most consistent and enduring targets” of conservative criticism (Jost et al., 2003, p. 351) since the right tends to perceive immigrants as a potential threat to their desire to maintain and justify the existing inequalities between White Americans and ethnic minorities/non-natives in terms of social inclusion and access to job opportunities (Hainmueller & Hopkins, 2014; Janoff-Bulman, 2009). The idea of promoting multiculturalism and inclusive immigration policies would therefore challenge conservatives’ expectations that the power status quo will remain the same.

According to Janoff-Bulman (2009), the difference in the way conservatives and liberals define group membership may explain why anti-immigration sentiments and attitudes tend to be a right-wing phenomenon. Conservatives stress “common social identity” to define the ingroup, and thus focus on immigrants’ outsider status, compared to liberals’ focus on the “common humanity” of immigrants (pp. 125–126). Considering the link between political conservatism and threat sensitivity (Oxley et al., 2008), as well as conservative hostility toward immigrants, the current study hypothesizes that:

H1: Political ideology is related to perceived immigration threat such that Republicans will perceive a larger threat than Democrats.

Conceptualizing threat perception as perceived difference between desires and perceived social reality as stated in false consensus research (e.g., Marks & Miller, 1987), one shall assume perceived threat from immigrants to be a response derived from the cognitive assessment of public opinion. Political communication scholars argue that both mass and interpersonal communication serve as vital information sources that individuals use to infer social reality. Citizens make inferences about public opinion climate about an issue based on their general impressions of news media coverage of that issue (Gunther et al., 2001), while their personal discussion networks of family, friends, and acquaintances are also known to play a significant role in shaping or altering their reality perceptions (Beck, 1991; Eveland & Shah, 2003). Although exposure to confirmatory information sources may reinforce one’s preexisting political beliefs, people may still perceive political media coverage in general to be unfavorable to their position or find themselves in political discussions with those who do not share their political views (Beck, 1991; Eveland & Shah, 2003). Given the potential relationship between perceived dissimilarity between the self and others and self-defense motives (Marks & Miller, 1987; Sherman et al., 1984), we suggest perceived political disagreement in communication at a mass or interpersonal level—that is, “perceptions of social disconfirmation”—to be a cognitive antecedent of perceived threat.

Through studies on media bias, one could theorize about the potential role that perceived media reality plays in evoking threat perceptions. Previous multi-wave panel studies suggest that one’s ideological leaning could significantly contribute to perceptions of media bias (Eveland & Shah, 2003; Huge & Glynn, 2010) regardless of actual media bias. Although partisans on both sides of an issue tend to perceive media coverage to be biased against their own side, some provide evidence that such tendency is more manifest among right-leaning citizens than Democrats (Eveland & Shah, 2003; Lee, 2005). This has especially been the case since the presidency of Donald Trump whose allies have routinely accused mainstream media of bias (Lee & Hosam, 2020).

If people were to infer public opinion through their perception/interpretation of mass media, thinking that others are also exposed to and affected by the mass media contents they consume (Gunther & Storey, 2003; Rojas, 2010), one could speculate that individuals who perceive the mainstream media (i.e., media that target the large public as opposed to alternative/niche media; Tsfati & Peri, 2006) to be biased against their political views would believe that public opinion in general is hostile to their views. While studies show mixed findings regarding whether endorsing a particular political ideology predicts the frequency of “cross-cutting political conversation,” or talking with political dissimilar others¹ (Barberá et al., 2015; Heatherly et al., 2017), we argue that conservatives would be more likely than liberals to believe that they engage in political conversation with those who hold differing views, perceiving themselves as opinion minorities by presuming the influence of mainstream media bias on others. A 2012 national survey (Heatherly et al., 2017) found that Republicans were more likely to be involved in cross-cutting political discussions than did Democrats, and the authors suggest that perceptions of liberal bias in media might have led Republicans to perceive political discussions as biased in favor of opposing views.

In this study, we expect a significant association between political ideology and cognitive assessment of social disconfirmation. More specifically, we hypothesize that conservatism positively relates to perceived mainstream media bias and perceived frequency of cross-cutting talk. And because people presume biased media contents to have an influence on others, we also expect a direct relationship between perception of media bias and cross-cutting talk.

H2: A politically conservative ideology is positively related to (a) perceived mainstream media bias and (b) perceived cross-cutting talk.

H3: Perceived media bias is positively related to perceived cross-cutting talk.

Based on the notion that feelings of threat may arise from perceiving minority status and hostile public sentiment (Marks & Miller, 1987; Morrison & Matthes, 2011), one should expect cognitive assessments of (unfavorable) communication environments to serve as determinants of perceived threat. Although people’s preexisting attitudes can prompt biased processing of information, perceived immigration threat has often been argued as an outcome of, rather than a reason for, the way people evaluate communication environments since much of what shapes or reinforces citizens’ attitudes toward immigration is derived from the mass media discourse about immigrants (Brader et al., 2008; McCabe et al., 2021) and the discussion with others about immigration (Lyons et al., 2011) rather than from direct contact with immigrants (Gravelle, 2016).

The mainstream media in the U.S. has evolved to reflect the increasingly multicultural nation by improving on-screen ethnic diversity (Bauder, 2015; Nielsen, 2018). The perceived prevalence of cross-cutting information in mainstream society would reinforce, not attenuate, conservatives’ prejudiced attitudes toward their outgroups. Kim (2019) found that conservatives’ prior opinions were strengthened through exposure to counter-attitudinal media, suggesting that perceived media bias and discussion disagreement may lead to self-defense motives among conservatives rather than accepting attitudes. Lee (2005) argues that the “liberal trends” reflected in mass media would lead conservatives to develop negative feelings and cynical attitudes toward societal changes and to see the media as their enemy (p. 56).

Several experiments have indeed found that exposure to information that reflects multiculturalism evokes negative attitudes toward racial minority among conservatives and White Americans, leading them to perceive threat to their group’s values and societal status (Craig & Richeson, 2014; Osborn et al., 2020). Such findings suggest a potential causal effect of cognition on perceived threat—that is, perceived discrepancy between desires and social reality (i.e., feelings of threat) may arise from cognitive evaluation of consumed information, at least in the case of conservatives regarding the issue of immigration. Considering political conservatives’ tendency to focus on immigrants’ outsider status when defining in-group/out-group membership (Janoff-Bulman, 2009; see also Proulx & Brandt, 2017), we argue that perceived disagreement in communication environments could act as “threat stimuli” that challenge conservative group norms. We therefore propose the following set of hypotheses to test the association between cognitive estimates of social reality and perceived threat, as well as the potential partial mediating role of perceived social disconfirmation in the relationship hypothesized in H1 (i.e., correlation between conservatism and perceived immigration threat):

H4: (a) Perceived mainstream media bias and (b) perceived cross-cutting talk are positively related to perceived immigration threat.

Immigration as Group Status Threat

Existing research on anti-immigration rhetoric in the U.S. indicates that the increasing ethnic diversity due to immigration poses group status threat to political conservatives (primarily White Americans) who feel that their higher status to immigrants is unstable or slipping (Major et al., 2018). The outgroup is perceived as a challenge to their established cultural norms, as well as a contender for limited resources such as jobs (Stephan et al., 2015), which have allowed scholars to identify two non-mutually exclusive (the frequently coexisting) subcategories of group status threat: symbolic and realistic threat.

Symbolic threats emerge from perceived group differences in cultural norms and standards. For conservatives, the American norms and standards tend to include traditional conceptions of membership in the national community, such as having birthright citizenship and being fluent in English, as well as the nation’s predominantly white demographic makeup (Hainmueller & Hopkins, 2014), all of which are perceived to be jeopardized by the influx of newcomers. Realistic threats refer to more tangible concerns and are known to stem from perceiving the outgroup as a threat to the physical/material well-being of the ingroup (Rios et al., 2018). Economic threat is a notable example of such threats as it reflects natives’ fear of losing jobs to immigrants or anxiety about immigrants’ impact on property demand and home prices (Coenders & Scheepers, 1998; Stephan et al., 2015). Perception of immigrants as a threat is often based on a zero-sum thinking in intergroup relations—social or economic status gains for the outgroup means status loss for the ingroup (Major et al., 2018).

Symbolic and realistic threats commonly exist side-by-side, collectively constituting conservatives’ perception of discrepancies between their desires and social reality. For example, many of those who supported Donald Trump during the 2016 presidential race saw the racial demographic shift as a threat to traditional American norms (symbolic) and worried that immigrants would take away jobs from American workers (realistic) (Major et al., 2018; Rios et al., 2018, p. 215). Scholars argue that evoking symbolic threats can prompt one to reflect upon economic issues, and vice versa (Brambilla & Butz, 2013), which leads us to take on a holistic approach to conceptualize perceived immigration threat as an experience that derives from cognitive assessment of information on a range of issues such as demographic diversity and economy.

The aforementioned subcategories of perceived threat are thus taken as a useful guideline for the current study in developing an exhaustive list of survey items for the threat construct. Based on the insights gained from prior literature on motivational determinants of false consensus (e.g., Krueger & Clement, 1997), we expect those who perceive immigrants as a threat to be more likely to overestimate consensus regarding their position on immigration.

H5: Perceived immigration threat is positively related to false consensus regarding immigration.

Our hypotheses suggest a serial relationship between political ideology, cognitive estimation, self-defense motive and false consensus. We additionally ask whether direct relationships exist between (1) cognition and false consensus and (2) political ideology and false consensus regarding immigration. From the cognitive availability perspective, perceived hostility of opinion climate would suggest to people that their own views are not as prevalent in the population, and thus would have a direct negative effect on false consensus, but since counter-attitudinal exposure was shown to bolster conservatives’ preexisting views (Kim, 2019), we ask whether perceived social disconfirmation positively relates to false consensus among conservatives. Additionally, despite the lack of empirical evidence regarding the link between political ideology and egocentric estimation of public opinion on immigration, some suggest a direct path from conservatism to false consensus (Stern, West, Jost, et al., 2014), citing intolerance of ambiguity and motivation to justify the normative legitimacy of own opinion as some of the characteristics of those who score high in conservative values (Eisner et al., 2020; Jost et al., 2003). Recognizing the inconclusive nature of such explanations in the context of false consensus and immigration, we propose the following research questions:

RQ1: Are (a) perceived mainstream media bias and (b) perceived cross-cutting talk related to false consensus regarding immigration?

RQ2: Is political ideology directly related to false consensus regarding immigration?

Figure 1 summarizes the theoretical model of our expectations and research questions.

Figure 1.

Conceptual path model predicting false consensus regarding immigration.

Method

Data

Data were collected through online survey panels administered by Qualtrics. A multistep sampling process was used to ensure the sample reflects the adult population in the U.S.: (1) subjects were randomly selected from the country’s online panel constructed based on geographic and demographic parameters; (2) subjects were then presented with profiling questions to create a final sample that reflects the demographic characteristics of the population. Potential respondents were given access to the survey through a web portal on the panel or received an email invitation explaining that the survey is for research purposes only and that members can unsubscribe at any time. They were informed about the length of the survey and the available incentives, but no specific details about the survey contents were included in the invitation.

The U.S. dataset was completed between February 27th and March 14th, 2020, and contains 510 complete responses (61% participation rate²). The sample reflects the U.S. adult population well in terms of gender (sample: 49.2% males; census data³: 49.2% males), age (sample: 18–34 years old 36%, 35–64 years old 46%, and 65 and over 18%; census: 18–34 years old 36%, 35–64 years old 46%, 65 and over 18%), race (sample: 76.6% identify as white; census: 76.3%), and level of education (sample: less than high school 9.2%, high school and/or some college 60.4%, completed college 30.4%; census: did not complete high school 10%, high school and/or some college 54%, completed college 36%). Cross-quotas of gender, age, and level of education revealed that the distributions closely reflected pre-defined quotas.

Measurement

Exogenous variables

Political ideology

Considering that party identification is a strong determinant of perceived ingroup threat and attitudes toward immigration (Gravelle, 2016), political ideology was assessed using a single-item self-identification scale ranging from 1 (strong Democrat) to 4 (Independent) to 7 (strong Republican), where higher numbers indicated more conservatism (M = 3.85, SD = 2.03). For the purpose of our analyses, we coded third party affiliation (8 = Other) as missing. Thirty-seven percent of the remaining 470 participants self-identified as either strong Democrat, Democrat, or lean Democrat (n = 173), 22% as independent (n = 102), and 41% identified as a strong, moderate, or lean supporter of the Republican Party (n = 195). The proportion of Republicans in our sample closely matched the finding from a national poll conducted during a similar time period (40%; Saad, 2020).

Endogenous variables

Cognitive assessment of social disconfirmation

Based on Ho et al. (2011), we measured participants’ perceptions of mainstream media bias using a single item that read, “On a scale from 0 to 5 (0 = strongly disagree, 5 = strongly agree), how much do you agree with the following statement? Mainstream news media tends to be biased” (M = 3.28, SD = 1.59). Following previous studies on perception of political disagreement (e.g., Barnidge, 2015), we assessed the level of perceived cross-cutting political talk by asking participants to indicate how often they talk about politics and current events with “people that don’t share [their] political views” on a scale from 0 (never) to 5 (frequently) (M = 1.70, SD = 1.62). For both items, higher scores indicated greater perception of disagreement in communication environment.

Perceived immigration threat

The measure of attitudes toward immigrants from the 2013 International Social Survey Program (ISSP) was modified and expanded to fit the purpose and scope of our study. On a 6-point Likert scale (0 = strongly disagree, 5 = strongly agree), participants were asked to indicate the degree to which they agree with the following statements, which were designed to reflect the previously mentioned subcategories of perceived immigration threat: (1) Immigrants increase crime rates; (2) American culture is generally undermined by immigrants; (3) In general, immigrants do not support the shared democratic values of Americans; (4) Immigrants take jobs away from people who were born in America; (5) Local culture is under threat because of the exposure to foreign cultures; (6) There is an increasing number of people in this country who speak little to no English; (7) I am worried that Americans will be outnumbered by immigrants; (8) My neighborhood has become unrecognizable because of immigrants; (9) Because of the influx of foreigners, the living cost in America has become higher. The items were averaged to create the final variable (M = 2.28, SD = 1.38, Cronbach’s α = .94). We performed a principal component analysis of the nine items (rotation method: Oblimin with Kaiser Normalization) and found that all items correlated with a single factor, which accounted for about 66% of the original variance.

False consensus regarding immigration

Based on previous false consensus research, we followed one of the most prevalent approaches to measuring consensus estimation bias, which is to calculate the difference between: (x) the subject’s estimate of the percentage of others who agree with her/his opinions and (y) the actual percentage of subjects in the survey who report holding such opinions (Marks & Miller, 1987; Suls et al., 1988). A positive x minus y difference is interpreted as overestimation, and a negative difference as underestimation of consensus. Higher absolute values indicate a greater magnitude of estimation error (Bauman & Geher, 2002). With this in mind, we took the following steps to compute the false consensus scores:

We first measured participants’ positive attitudes toward immigration using the two 6-point scale items (0 = strongly disagree, 5 = strongly agree) from the ISSP survey that read, “Immigrants improve society by bringing new ideas and cultures” and “Immigrants are generally good for the American economy,” which measured one’s positive perceptions regarding immigrants’ cultural and economic impact, respectively. The two items were averaged to create the variable concerning positive attitudes (M = 2.74, SD = 1.45, Cronbach’s α based on standardized items = .82; Eisinga et al., 2013).

We then subtracted perceived threat (i.e., negative attitude) scores from the positive attitude scores to produce “total attitude scores” for all participants. A minus total attitude value indicated negative attitudes toward immigrants, while zero indicated neutral attitudes. Fifty-four percent of participants reported having positive attitudes toward immigrants, 7% had neutral attitudes, and 39% had negative total attitude scores. We will refer to these percentages as “actual-consensus scores,” which were assigned to all participants according to their total attitude scores.

To measure participants’ estimation of others who hold the same position as theirs, our survey used an open-ended question that asked, “In your opinion, what % of the American population agrees with your position on immigration?” which required participants to enter a number from 0 to 100.

And finally, we subtracted the actual-consensus scores from the estimation scores to create our final false consensus variable (M = 6.00, SD = 27.42)⁴—positive scores were interpreted as overestimation, and negative scores as underestimation. For example, if a subject with a positive total attitude score (i.e., actual-consensus score = 54) estimated that 74% of the American population agrees with her/his position, the false consensus score for that subject was a plus 20. The false consensus variable was entered as a continuous measure into our model, higher scores reflecting greater overestimation of consensus.

Control variables

Demographics

We treated the following demographic variables as controls in our analyses: gender (female = 50.8%; male = 49.2%), age (M = 46.08 years, SD = 17.91), education, and income. The average education level was high school completion (M = 4.85, SD = 1.26) on a 7-point scale (1 = none, 7 = graduate degree). We used an 8-point scale to measure monthly household income in US dollars (M = 4.28, Mdn = $3,001–$4,000, SD = 2.48).

Traditional media use

Given its close association with perceived opinion climate (Gunther et al., 2001) and false consensus bias (e.g., Wojcieszak, 2011), frequency of traditional media use was controlled for in our analyses. The survey asked participants to indicate on a 6-point scale (0 = never, 5 = frequently) how often they read, watch or listen to: (1) national daily newspapers; (2) national television news; (3) foreign news outlets; (4) national entertainment products (e.g., television, films, music). The items were averaged to create an index of traditional media use (M = 2.50, SD = 1.55, Cronbach’s α = .72).

Social media use

On a scale from 0 (never) to 5 (frequently), participants indicated how often they read, watch or listen to “social media news (e.g., Facebook/Twitter)” (M = 2.92, SD = 1.94). The variable was held constant throughout the analyses since the cognitive explanations to false consensus mainly assume that estimation bias could arise from selective exposure to consonant information (Marks & Miller, 1987), a phenomenon that has frequently been discussed in research on social media use (e.g., Bakshy et al., 2015).

Political interest

Participants were asked to indicate how interested they are in politics on a 6-point scale (0 = not at all, 5 = a lot) (M = 2.72, SD = 1.70). The variable was held constant throughout the analyses since it has been found to be related to perceived media bias and public opinion perceptions (Schulz et al., 2020).

Before testing our hypotheses and research questions, we employed Pearson’s correlation analyses to examine relationships between the variables of interest (see Table 1). None of the correlation coefficients suggested multicollinearity (Tabachnick & Fidell, 2007). All significant pairwise correlations were positive, including that between perceived cross-cutting political talk and false consensus (β = .23, p < .01). Perception of mainstream media bias was not associated with false consensus regarding the issue of immigration (p = .13).

Table 1.

Zero-Order Correlations Between Variables of Interest.

	(1)	(2)	(3)	(4)	(5)
Political ideology (1)	1.00
Media bias (2)	.22***	1.00
Cross-cutting talk (3)	.10*	.22***	1.00
Perceived threat (4)	.33***	.26***	.35***	1.00
False consensus (5)	.16***	.09	.12**	.41***	1.00

Note. N = 470.

p < .05. **p < .01. ***p < .001.

Results

Path analysis in STATA 16.0 was performed to test the hypothesized model linking political ideology to perception of social disconfirmation, perceived immigration threat and false consensus. Results indicated that the model was a strong fit for the data according to the following fit measures: the chi-square goodness-of-fit index, the comparative fit index (CFI), the root mean square error of approximation (RMSEA), and the standardized root mean square residual (SRMR) (see Table 2). All lines shown in Figure 2 indicate significant paths. Standardized path coefficients are reported.

Table 2.

Goodness-of-Fit Measures for Path Analysis.

Measure	Statistic	p Value	df
χ²	2.634	.621	4
CFI	1.000	—	—
RMSEA	0.000	.916	—
SRMR	0.006	—	—

Note. CFI = comparative fit index; RMSEA = root mean square error of approximation; SRMR = standardized root mean square residual.

Figure 2.

Standardized path coefficients (N = 470).

Our analysis revealed significant paths from political ideology to the likelihood of perceiving immigrants as a threat and perceptions of disagreement in communication environment. We report a direct positive association between conservatism and perceived threat (H1: γ = .24, p < .001), as well as the positive paths from ideology to perceived mainstream media bias (H2a: γ = .23, p < .001) and perceived cross-cutting political talk (H2b: γ = .11, p < .01). Our findings did not support H3, which predicted that perceiving media bias would be significantly related to one’s perceived frequency of cross-cutting conversations (p = .11). Both variables regarding the cognitive assessment of social disconfirmation were positively associated with perception of threat (H4a: β_MediaBias = .16, p < .001; H4b: β_{CrossCuttingTalk} = .31, p < .001), acting as partial mediators in the relationship between ideology and perceived threat from immigrants. The total (i.e., direct and indirect) relationship between political ideology and perception of threat was significant (γ = .31, p < .001). The model accounted for about 30% of the residual variance related to perceived immigration threat.

As expected, our path model indicated a significant positive path from perceived threat to false consensus regarding immigration (H5: β = .37, p < .001). Overestimation of consensus was not significantly correlated with perceived media bias (RQ1a: p = .59) or cross-cutting talk (RQ1b: p = .06), although the two communication variables were indirectly associated with estimation bias (β_MediaBias = .06, p < .01; β_{CrossCuttingTalk} = .11, p < .001) through perceived threat. The direct path from political ideology to false consensus effect was insignificant in our model (RQ1a: p = .58), but political conservatism was shown to have an indirect effect on consensus bias (γ = .12, p < .001). The integrated model explained about 20% of the residual variance for false consensus regarding immigration.^5,6 Table 3 presents the standardized path coefficients for the trimmed path model after removing the nonsignificant paths.

Table 3.

Standardized Direct, Indirect, and Total Effects of the Trimmed Path Model (N = 470).

	Outcome	Explanatory variable	β (SE)	95% CI	p
Direct effect	Perceived media bias	Political ideology	.23 (0.03)	0.10, 0.24	<.001
	Perceived CCT	Political ideology	.11 (0.03)	0.02, 0.14	.003
	Perceived threat	Political ideology	.24 (0.03)	0.11, 0.22	<.001
		Perceived media bias	.16 (0.04)	0.06, 0.21	<.001
		Perceived CCT	.31 (0.04)	0.18, 0.34	<.001
	False consensus	Perceived threat	.37 (0.90)	5.29, 8.83	<.001
Indirect effect	Perceived threat	Political ideology	.07 (0.01)	0.02, 0.07	<.001
	False consensus	Political ideology	.12 (0.28)	0.97, 2.07	<.001
		Perceived media bias	.06 (0.31)	0.44, 1.64	.001
		Perceived CCT	.11 (0.37)	1.13, 2.58	<.001
Total effect	Perceived media bias	Political ideology	.23 (0.03)	0.10, 0.24	<.001
	Perceived CCT	Political ideology	.11 (0.03)	0.03, 0.15	.006
	Perceived threat	Political ideology	.31 (0.03)	0.16, 0.27	<.001
		Perceived media bias	.16 (0.04)	0.07, 0.22	<.001
		Perceived CCT	.31 (0.04)	0.18, 0.34	<.001
	False consensus	Political ideology	.12 (0.60)	0.76, 3.10	.001
		Perceived media bias	.06 (0.31)	0.44, 1.64	.001
		Perceived CCT	.11 (0.37)	1.13, 2.58	<.001
		Perceived threat	.37 (0.90)	5.29, 8.83	<.001

Note. SE = Standard error; CI = Confidence interval; CCT = Cross-cutting talk.

We tested an alternative model that proposed perceived threat as an antecedent of political ideology, given that motivation to avoid threat and uncertainty (e.g., perceived discrepancy between desires and reality) could serve as a reason for adopting conservative beliefs (Jost et al., 2003). Earlier studies have also shown that anti-immigrant sentiments could lead to support for restrictive immigration policies and the political party that advocates them (Major et al., 2018; Watson & Riffe, 2013). The chi-square goodness-of-fit test and the RMSEA value indicated that the Cognition-Threat-Ideology-Overestimation model provided a bad fit to the data (χ²(4) = 14.530, p = .002; RMSEA = 0.090, p = .059). Although the direction of the path from cognition to perceived threat proposed in our original model is derived from the notion that people’s attitudes toward immigration largely depend on media/interpersonal discourse about the issue, we ran a model in which perceived threat was established as the sole exogenous variable (i.e., Threat-Cognition-Ideology-Overestimation) since earlier research indicated that strength of issue stance could determine perceptions of public opinion hostility (e.g., issue of gun rights; Chung et al., 2015) and the political party/candidate that one decides to endorse (Fournier et al., 2003). The model indicated good fit for our data (χ²(5) = 3.438, p = .441; CFI = 0.999; RMSEA = 0.001, p = .716; SRMR = 0.006) though it did not provide a statistically better fit than our main model. Perceived threat was identified as the only predictor of false consensus in the model, explaining less than 20% of the residual variance in overestimation of consensus.

To further investigate our findings from the main model, we ran exploratory post hoc analyses using the conservatives-only sample (n = 195) to examine whether any of the variables entered as controls in our previous analyses affect the theorized paths when included in the model. The zero-order correlation analyses showed that false consensus regarding immigration (M = 11.77, SD = 28.31) was positively associated with social media use (M = 3.09, SD = 1.96) (β = .18, p < .05) and political interest (M = 2.92, SD = 1.62) (β = .15, p < .05). In the subsequent path analyses, we substituted political ideology, which was entered as the exogenous variable in our main mediation model, for either frequency of social media use for news or political interest. The social media model indicated non-significant paths from social media use frequency to perceived mainstream media bias (p = .26), perceived cross-cutting talk (p = .09), and perceived threat (p = .051). The model as a whole was shown to be a bad fitting model according to the chi-square goodness-of-fit statistics and RMSEA (χ²(5) = 4.316, p = .038; RMSEA = 0.130, p = .087).

On the other hand, our political interest model, controlling for demographics and traditional, social media use, showed good fit for the data (χ²(5) = 4.114, p = .533; CFI = 1.000; RMSEA = 0.000, p = .753; SRMR = 0.014) (see Figure 3 for standardized path coefficients—only significant paths are shown—and Table 4 for a summary of the direct, indirect, and total effects). Political interest was positively associated with both perceived mainstream media bias and perceived cross-cutting political discussion. Our analysis revealed significant positive paths from perceptions of political disagreement to perceived threat. Political interest was not directly associated with perceived threat, as it was shown to have indirect effect on perceived threat through variables concerning perceived social disconfirmation. The path from perceived immigration threat to overestimation of consensus was significant. The predictors explained 24% of the residual variance in perceived threat, and about 17% of the variance in false consensus among conservative samples.

Figure 3.

Standardized path coefficients for conservatives (n = 195).

Table 4.

Standardized Direct, Indirect, and Total Effects of the Trimmed Path Model for Conservatives (n = 195).

	Outcome	Explanatory variable	β (SE)	95% CI	p
Direct effect	Perceived media bias	Political interest	.17 (0.07)	0.08, 0.30	.038
	Perceived CCT	Political interest	.23 (0.07)	0.08, 0.37	.003
	Perceived threat	Perceived media bias	.16 (0.05)	0.02, 0.22	.019
	Perceived threat	Perceived CCT	.39 (0.05)	0.18, 0.38	<.001
	False consensus	Perceived threat	.37 (1.61)	4.93, 11.22	<.001
Indirect effect	Perceived threat	Political interest	.11 (0.03)	0.03, 0.13	.002
	False consensus	Political interest	.04 (0.25)	0.17, 1.15	.008
		Perceived media bias	.05 (0.46)	0.08, 1.90	.033
		Perceived CCT	.13 (0.62)	1.04, 3.46	<.001
Total effect	Perceived media bias	Political interest	.17 (0.07)	0.01, 0.30	.038
	Perceived CCT	Political interest	.23 (0.07)	0.08, 0.37	.002
	Perceived threat	Political interest	.11 (0.03)	0.03, 0.13	.002
		Perceived media bias	.16 (0.05)	0.02, 0.22	.019
		Perceived CCT	.39 (0.05)	0.18, 0.38	<.001
	False consensus	Political interest	.04 (0.60)	0.17, 1.15	.008
		Perceived media bias	.05 (0.46)	0.08, 1.90	.033
		Perceived CCT	.13 (0.62)	1.04, 3.46	<.001
		Perceived threat	.37 (1.61)	4.93, 11.22	<.001

Note. SE = Standard error; CI = Confidence interval; CCT = Cross-cutting talk.

Discussion

The present study was an attempt to provide motivational explanations to false consensus, a phenomenon that has frequently been described by communication scholars as an outcome of cognitive processing of available information. We specifically focused on perceived threat as a motivational determinant of consensus estimation bias, which is known to result from one’s awareness of discrepancies between desires and perceived reality in contemporary communication environment that expose citizens to diverse political perspectives. Combining the notion of false consensus as a self-defense mechanism to minimize perceived threat with the existing knowledge that links political conservatism and greater threat sensitivity, we used a nationally representative sample of U.S. adults to examine the relationship between political ideology, cognition, perceived threat, and false consensus regarding one of the most divisive political issues facing America today.

Our main serial mediation model revealed a significant positive path from political conservatism to the likelihood of perceiving immigrants as a threat, which was partially mediated by perceptions of social disconfirmation, that is, one’s subjective assessments of disagreements in mass-mediated and interpersonal communication environment. We found that such cognitive factors were not directly associated with self-serving estimates of consensus, but instead were fully mediated by perceived threat, which was found to be the strongest and the only construct that was significantly related with false consensus effect. Evidence derived from cross-sectional data will not account for causal explanations, but the alternative mediation models tested in our research lend weight to the view that feelings of threat might be triggered when the mainstream media and political discussion partners are considered to be counter-attitudinal to one’s position, and that perceiving threat in turn may motivate people to overestimate their agreement with others. Our findings propose that political ideology could serve as the very antecedent of the hypothesized relationships between cognition, motivation and false consensus, and that the sequential approach to linking ideological leaning with consensus overestimation might be more applicable to conservatives, at least when the discussion centers on the issue of immigration.

The results obtained from the conservatives-only model support the idea that political interest may increase perceived media bias (Schulz et al., 2020), and further suggest a positive relationship between political interest and perceived cross-cutting talk among those who support the Republican Party. The significant indirect paths from political interest to perceived threat and consensus overestimation provide interesting insights given that high political interest has been argued to “[benefit] society and individuals” in a democracy (for review, see Lupia & Philpot, 2005), but the findings are interpreted as theoretical questions rather than answers regarding the concept of political interest from our study’s perspective. Though it seems valid to use political interest as a control variable when it is assumed to be a personality trait (e.g., Caspi et al., 2005), if the concept were to be defined as “willingness to pay attention to political phenomena” (Lupia & Philpot, 2005, p. 1122), one could advocate a cognitive-motivational approach to political interest to elaborate on its potential relation to perceived threat and false consensus effect.

Having said that, perhaps one of the most noticeable limitations in this research relates to the self-report measures that we used to test our expectations and inquiries. We acknowledge that our single-item measures may not have fully captured the possibly multidimensional aspects of perceived mainstream media bias (e.g., differences across media sources) and cross-cutting political talk (e.g., formal vs. informal talk), which might serve as a possible explanation for the insignificant result regarding the relationship predicted in H3. And although we believe that various types of news/entertainment media contents and different political conversation topics have the potential to elicit thoughts about immigrants’ impact on culture and economy (e.g., anything that reminds people of the current multicultural trends), one could claim that the survey items could have been tailored to specifically focus on the issue of immigration to better establish the link between cognition and perceived threat. Additionally, while political ideology is often assessed via party identification measures in the context of immigration in a two-party system, the current study cannot generalize its results to conservative ideology at large, which would require the use and combination of various ideological measures—including, but not limited to, the general left-right scale and measures of authoritarianism and attitudes toward multiculturalism. And finally, although the directions of many of the paths hypothesized in our model are based on earlier experimental findings and panel data evidence (e.g., Eveland & Shah, 2003; Osborn et al., 2020), our analysis calls for experimental/longitudinal approaches to test the causal ordering of the key variables used in this study, which are not yet available in the literature.

Nonetheless, we believe that the present research makes important contributions to the existing body of political communication literature on false consensus by demonstrating that: (1) motivational factors, in the case of threats to group status, might fully mediate the relationship between cognition and people’s overestimation of population consensus for their position, and that (2) the level of perceived threat and the likelihood of overestimating consensus regarding a partisan issue may relate more strongly with supporters of a particular political party. These insights should prompt further scholarly investigation into the generalizability of our findings to threats regarding different political issues including the ones to which Democrats are known to be more sensitive (e.g., climate change; Hamilton, 2011), issues that are perceived to be “owned” by the Democratic Party (e.g., civil rights; Bélanger & Meguid, 2008), as well as those that are relatively less polarized in American politics (e.g., affordability of college education; Pew Research Center, 2019). From the motivational approach to false consensus, one would expect consensus overestimation to serve as a self-defense mechanism against perceived threat regardless of issues being discussed, but the question of whether or not the significant positive relationship between perceived social disconfirmation and feelings of threat is specific to issues that pose external threat to conservatives remains unanswered. Additionally, samples from multiple countries would enable future studies to evaluate the cross-national replicability of our findings. One could argue that our main thesis regarding the path from cognition to false consensus through motivation would be largely applicable to countries with a two-party system and a highly polarized media landscape (e.g., South Korea), although it remains an interesting question as to whether the demonstrated relationship between party identification and other variables of interest would hold for countries with a multiparty or a dominant-party system. It is plausible that in countries with lower levels of political polarization, ideology might be less predictive of bias perceptions and in turn less predictive of perceived threat and false consensus through it. In terms of multiparty systems, parties in these could also align under bipolar logics (Schmitt-Beck & Partheymüller, 2016), so the potential for biased perceptions is there at least for parties holding more extreme positions (Garry, 2007).

We also believe that our discussion on the partial mediating role of cognition in the link between Republican Party support and perceived immigration threat can be further narrowed down to address different components of perceived media bias such as perceived balance and credibility, as well as those of cross-cutting political conversation (e.g., weak vs. strong tie discussion partners). In terms of attitudes toward immigration, future studies should examine whether immigrants of different origins pose different types of threat to the natives (e.g., Asian Americans as a source of economic threat; Rios et al., 2018) to further explore the underlying dimensions of perceived immigration threat that may be mutually exclusive to one another depending on the race/ethnicity of immigrants. Along the line, we anticipate future works to address “perceived target of threat,” that is, to develop survey measures that distinguish egocentric concerns (e.g., personal financial situation) from group-level or sociotropic concerns (e.g., national economy) regarding immigration (Lockerbie, 2006) to see if such categorization yields meaningful variations in the level of false consensus.

The insights gained from our mediation model should provide a starting point for future experiments on motivation and false consensus, but the variables explored in this study are by no means an exhaustive list of antecedents of perceived threat and self-serving estimate of public opinion. Possible determinants of group status threat among conservatives may include perceived public support for the Democratic Party and inclusive immigration policies, and we speculate that one’s “need to belong” may serve as an additional motivational factor that contributes to false consensus effect (Morrison & Matthes, 2011). Lastly, false consensus as a psychological phenomenon should elicit scholarly inquiry about its behavioral consequences. In addition to higher willingness for political participation in the form of collective activism (Stern, West, Jost, et al., 2014, pp. 1162–1163), we argue that stronger perceptions of consensus might also lead one to be more vocal about supporting a controversial public policy or a political figure who endorses it, since overestimating the number of likeminded others could mitigate social pressure to openly express one’s opinions.

While the current research could be understood as an effort to reinforce the notion of partisan motivated reasoning and communication theories that remind people of the irreconcilable gap between the two ideological poles, it is in fact hoped that the insights gained here help alter the way one thinks about those on the other side of the political spectrum. Exaggerating public support for own political attitudes may not necessarily be an indication of living in an echo chamber of irrationality, but could rather be a sign of sensitivity to threat. And if self-protection against threat were to be regarded as a basic human motive, this study shall be considered a small step toward tolerance.

Footnotes

Declaration of Conflicting Interests

The authors declared no potential conflicts of interest with respect to the research, authorship, and/or publication of this article.

Funding

The author(s) disclosed receipt of the following financial support for the research, authorship, and/or publication of this article: This work was supported by University of Wisconsin Foundation: [Grant Number AAA2223].

ORCID iD

Hyungjin Gill

Notes

Author Biographies

Hyungjin Gill is a doctoral candidate at the University of Wisconsin-Madison School of Journalism and Mass Communication. His research focuses on examining how digital social networks affect civic engagement and party identification, and aims at identifying mobile user experience that shapes attitudes toward political information.

Hernando Rojas is Professor of Journalism and Mass Communication and Helen Firstbrook Franklin Chair at the University of Wisconsin-Madison. His scholarship focuses on political communication, in particular examining: (a) the deployment of new communication technologies for social mobilization in a variety of contexts; (b) the influence of audience perceptions of media (and audience perceptions of media effects) on both public opinion and the structure of the public sphere; and (c) the conditions under which media support democratic governance.

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