Source Cues, Partisan Identities, and Political Value ...users.polisci.umn.edu/~pgoren/Goren AJPS 09.pdf · Source Cues, Partisan Identities, and Political Value Expression Paul Goren

Source Cues, Partisan Identities, and Political ValueExpression

Paul Goren University of MinnesotaChristopher M. Federico University of MinnesotaMiki Caul Kittilson Arizona State University

This article examines the conditions under which partisan identities shape the positions people express on four politicalvalues: equal opportunity, self-reliance, moral traditionalism, and moral tolerance. The theoretical framework posits that(1) party source cues activate latent partisan biases in the minds of citizens, which in turn affect the degree to whichindividuals express support for these values; (2) out-party cues are more powerful motivators of value expression thanin-party cues; (3) value shifts are more pronounced when liberal-conservative identities reinforce partisan sentiments; and(4) partisan cues promote horizontal constraint among these values. These hypotheses are tested using data from a set ofexperiments appearing on a novel national survey. The empirical results generally support these theoretical expectations.

Party identification represents the most stable andinfluential political predisposition in the belief sys-tems of ordinary citizens. As Campbell and his

colleagues put it many years ago, “party has a profoundinfluence across the full range of political objects to whichthe individual voter responds” (1960, 128). The influ-ence of party over short-term political objects is well es-tablished. Accumulated research demonstrates that partyshapes judgments about the economy, presidential ap-proval, candidate evaluations, voter choice, and policypreferences (Bartels 2002; Green and Palmquist 1990;Goren 2002; Layman and Carsey 2002). Does party alsoshape evaluations of more abstract and enduring objectssuch as political values? And do party cues promote con-straint among political values? If the answer to each ofthese questions is yes, then the influence of partisanshipover political judgment may be even more profound thanCampbell, Converse, Miller, and Stokes imagined.

This article seeks to strengthen the case for parti-san influence by examining whether partisan source cues(1) lead party identifiers to modify their expressions ofsupport for a series of political values and (2) enhance

Paul Goren is Associate Professor of Political Science, University of Minnesota, Minneapolis, MN 55455 ([email protected]). Christopher M.Federico is Associate Professor of Psychology and Political Science, University of Minnesota, Minneapolis, MN 55455 ([email protected]).Miki Caul Kittilson is Associate Professor of Political Science, Arizona State University, Tempe, AZ 85287 ([email protected]).

We thank the Institute for Social Science Research at Arizona State University for funding this research and Hooi Hong Khor and PamelaHunter for administering our survey. In addition, we thank Robert Erikson, Tim Johnson, Joanne Miller, the four anonymous reviewers,and Marianne Stewart for their comments and suggestions on how to improve earlier versions of the manuscript. All remaining errors areour own.

constraint between these values in mass belief systems.We make two fundamental theoretical claims. First, weidentify conditions under which party cues should affectsupport for political values. Our theory holds that partysource cues activate latent partisan biases in the mindsof individuals, which in turn affect their willingness toexpress support for several core political values–equal op-portunity, self-reliance, moral traditionalism, and moraltolerance; that out-party cues manifest stronger influencethan in-party cues; and that these effects are most pro-nounced among ideologically congruent partisans. Sec-ond, our theory maintains that party cues augment valueconstraint. In this way, value structure can be seen as apartial function of partisan cue taking.

We utilize data from a series of experiments em-bedded within a national survey to demonstrate thatwhen given party source cues, Democrats and Repub-licans move farther apart in their expressions of supportfor political values. We further show that opposition-party cues produce larger effects than in-party cues andthat value polarization is especially pronounced amongideologically pure partisans. Finally, we find that the

American Journal of Political Science, Vol. 53, No. 4, October 2009, Pp. 805–820

C©2009, Midwest Political Science Association ISSN 0092-5853

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806 PAUL GOREN, CHRISTOPHER M. FEDERICO, AND MIKI CAUL KITTILSON

influence of party extends to leading individuals to holdmore tightly constrained value systems. Put simply, partyidentification shapes the positions citizens take on dis-crete political values and promotes constraint in politicalvalue systems.

The Nature of Partisan Influence

We begin by defining the key concepts and then explainwhy cues pertaining to party identification can be ex-pected to shape core political values individually and col-lectively. Party identification represents an enduring senseof psychological attachment to a political party. Its im-portance in models of preference formation and politicalbehavior arises from findings that partisan identities arethe most stable predisposition in political belief systems;that they color a wide range of short-term political per-ceptions; and that they lie beyond the influence of mostshort-term perceptions (Campbell et al. 1960).1 Core po-litical values are abstract normative beliefs about desirableend-states or modes of conduct that operate in the po-litical domain (Feldman 1988; Rokeach 1973; Schwartz1994). They, too, are quite stable and guide preferenceson short-term political controversies and the issues of theday.

The first theoretical question to address is how partyidentification shapes political perception and judgment.Political parties serve as reference groups for citizens byproviding them with a simple evaluative basis for render-ing judgments about political communications (Camp-bell et al. 1960; Jacoby 1988). Although Democrats andRepublicans in the electorate may sometimes sponta-neously process political information in a partisan fash-ion, they are more likely to do so when primed bythe parties and their leaders to think in partisan terms(Zaller 1992). When someone hears a recognizable parti-san source advocating some position, her partisan lean-ings are activated, which in turn lead her to evaluate themessage through a partisan lens. If the cue giver andrecipient share a party label, the latter will trust the for-mer and accept the message without reflecting much onmessage content. But if the cue giver and recipient lieacross the partisan divide, the recipient will mistrust thesource and reject the message, again without much re-flection. In sum, when it comes to political communi-cations most people do not analyze systematically themessages they encounter. Instead, they usually turn tosimple heuristics, such as whether or not they like or trust

1Some issue preferences affect partisan loyalties among narrowlydefined issue publics. Even here, the effects of party eclipse thoseof issues (Carsey and Layman 2006).

the source, when evaluating a message (Carmines andKuklinski 1990; Mondak 1993).

The proposition that party sources lead partisans tomodify their positions on short-term perceptions like thestate of the economy or presidential popularity rests on afirm empirical foundation (Bartels 2002). We posit thatpolitical values are susceptible to partisan influence aswell. Thus, the second theoretical question we must ad-dress concerns why partisan bias should influence po-litical values, which, conventional theorizing holds, aremore durable and resistant to change than short-termperceptions. Put simply, we believe that partisan identi-ties are more central and enduring elements in mass be-lief systems than political values.2 Partisan identities aregrounded in group attachments, which are easier to de-velop and understand than normative beliefs about whatshould be done in a given policy domain. In light of this,we suspect that partisan identities precede political beliefin the developmental sequence. Green, Palmquist, andSchickler capture this sentiment well: “When people feela sense of belonging to a given social group, they absorbthe doctrinal positions that the group advocates” (2002,4). Just as religious adherents often accept matters of be-lief and faith without much critical thought, so too shouldpartisan adherents accept party doctrine on fundamentalpolitical questions.

Recent evidence is consistent with this point. Goren(2005) asks whether party ties shape political values orvice versa. He compares the partisan-influence hypoth-esis, which holds that group identities shape value posi-tions, with the values-influence hypothesis, which positsthat values shape partisan loyalties. These hypotheses aretested by applying structural equation modeling tech-niques to two- and three-wave NES panel data to estimatea series of cross-lagged models between party identifica-tion and equal opportunity, limited government, fam-ily values, and moral tolerance. The results suggest thatparty identification is more temporally stable than thesevalues and that it dynamically constrains values, withoutbeing simultaneously constrained by them. In addition,research by McCann (1997) uses NES panel data to showthat beliefs about equality and traditional morality aredynamically constrained by candidate evaluations, but donot manifest comparable influence over candidate eval-uations. Given that candidate evaluations are groundedheavily in partisan biases, McCann’s results yield indirectevidence that partisanship shapes value positions. In con-junction, these two studies provide solid justification for

2To be clear, we are talking about values that have become politi-cized. We make no claims regarding less political, more personalvalues.

CUES, PARTY, AND VALUES 807

positing that party identification has temporal priorityover, and thus, the ability to shape, some political values.

To be clear, we are not claiming that core politicalvalues are akin to other short-term perceptions. We be-lieve, like many others, that political values are centralelements in mass belief systems. Nor do we expect thatpartisan influence will produce wholesale value change.Instead, partisan forces should affect values at the mar-gins, essentially leading identifiers to become a little bitmore or less enamored of a given value.

To summarize our first theoretical claim: since par-tisan identities are more basic orientations than politi-cal values and partisan biases are activated easily in thepresence of partisan source cues, we anticipate that theexpression of political values will move in response topartisan appeals. This theoretical framework leads to thefollowing pair of expectations. Our first hypothesis isthat the relationship between individual partisan iden-tities and positions on political values will become morepronounced in the presence of party source cues. Our sec-ond hypothesis is that when political values are endorsedby Democrats, support will rise among Democratic iden-tifiers and fall among Republicans. Similarly, values ac-companied by Republican endorsements should see a risein support among GOP partisans and declining supportamong Democrats.

Two refinements are necessary to establish someboundary conditions for our theory. First, there aregrounds for suspecting that source cues will have asym-metrical effects on value expression. Specifically, partycues should produce larger changes in value supportamong opposition partisans compared to like-mindedpartisans. In other words, a type of negativity bias shouldbe present. The superior power of negative informationover positive information is firmly established in psy-chological literatures (for a review, see Baumeister et al.2001). Although several explanations exist as to why neg-ative information is more powerful, the most relevant, inour view, is the “figure-ground” explanation. This per-spective holds that negative information is more conse-quential for information processing and judgment thanequally extreme, positive information because it standsout against a positive background (Lau 1985).

Since political values are quite popular, the percep-tual background surrounding them is positive. Hence,when perceivers encounter a value message with an out-party endorsement, the negative valence of the out-partycue will contrast with the positive valence of the value.However, for perceivers who share a party label with asource, the entire message is positive. The net result is thatout-partisans, who confront some negative information,should move more than in-partisans, who encounter en-

tirely positive information. To illustrate, Democratic en-dorsement of equal opportunity should diminish Repub-lican support more than it enhances Democratic support.This leads to our third hypothesis. When a political valueis sanctioned by a political party, the drop in expressedsupport for the value among opposition partisans will begreater than the rise in expressed support among like-minded partisans.3

Our next theoretical refinement centers on the roleliberal-conservative identification likely plays in the pro-cess. We define this construct as a sense of psychologi-cal attachment to an ideological label based on feelingsabout liberal or conservative groups, ideas, or symbols(Conover and Feldman 1981). We think that the influenceof the party source cue will be stronger among ideolog-ically congruent partisans (i.e., liberal Democrats) thanwhen ideologically congruent and incongruent partisansare grouped together (i.e., liberal and moderate and con-servative Democrats).

Why should this be so? Imagine a pair of Democratsidentical in all respects save that the first is conserva-tive, the second liberal. Upon hearing that conservativeRepublicans support the principle of self-reliance, theirreactions may differ. The liberal Democrat confronts rein-forcing pressures—one partisan, one ideological—to coolher enthusiasm for self-reliance. In contrast, the conser-vative Democrat is cross-pressured: partisanship dictatesless enthusiasm, ideology more. It seems plausible that theliberal Democrat will express less support for self-reliancethan the conservative Democrat. When we compare re-sponses for liberal Democrats alone to those of liberalsand moderates and conservatives inclusive, the ideologi-cally “pure” should prove more responsive to source cuesthan the ideologically “mixed.” Thus, our fourth hypoth-esis predicts greater shifts in value endorsement amongideologically congruent partisans versus the ideologicallypure and cross-pressured.

To repeat, the first major prediction our theory makesis that party cues will condition the impact partisan loyal-ties have on political values under varying circumstances.In addition to shaping individual value positions, we ex-pect party cues will influence how discrete values areorganized into broader value systems. More succinctly,our final hypothesis predicts that party cues will enhance

3Note that we are looking for negativity bias in situations whereit should be most likely to obtain. For instance, Democratic en-dorsement of equal opportunity should produce stronger negativereactions for Republicans than positive reactions for Democrats.For reasons to be discussed below, we are not looking at whathappens if Republicans praise equality. In this case, one might ex-pect GOP support for equal opportunity to rise while Democraticsupport holds steady. That is to say, negativity bias would not bepresent.


constraint between political values. In his classic essay,Converse defines constraint “to mean the success wewould have in predicting, given initial knowledge that anindividual holds a specified attitude, that he holds certainfurther ideas and attitudes” (1964, 207). To illustrate forthe case of political values, imagine knowing someonewho favors equal opportunity. Armed with this knowl-edge, one might further predict she favors moral toler-ance. If she holds these two positions, we can concludethat her values are horizontally constrained. Constraint,in other words, denotes left-right cognitive organization.

Why should party cues enhance value constraint?When individuals hear that Democrats favor one set ofvalues and Republicans another, they should modify theirexpressed value positions for the reasons discussed above.This is the first-order effect of partisanship. Because ofthis, value positions will become more polarized, the re-sult of which will be stronger left-right constraint amongall the values. This is the second-order effect. In this man-ner, party cues help people organize their political val-ues more tightly, just as they promote attitude constrainton specific issues (Layman and Carsey 2002; Tomz andSniderman 2005). Insofar as this is the case, we can inferthat the influence of partisanship extends beyond simplyshaping value expressions.

The Partisan Influence Experiments

We tested our hypotheses by carrying out survey experi-ments that manipulated whether or not respondents re-ceived source cues when answering a series of questionsabout political values. The survey was administered us-ing random digit dialing techniques to the U.S. votingage population in the 48 contiguous states and was con-ducted from February 11 to April 26, 2006, by the Institutefor Social Science Research at Arizona State University.A total of 1,201 interviews were completed. We restrictour attention to respondents in three of the four condi-tions as described below, which yields a sample of 899. 4

Given our interest in the influence of partisanship on ex-pressed value support, we emphasize the internal validityof our results over their external validity throughout theanalyses.5

4The fourth condition provides liberal/conservative source cueswithout reference to the parties. Given our interest in partisaninfluence, we do not use respondents from this group.

5Our sample is representative of the broader U.S. public in manyrespects, though it is not a perfect reflection of the populationdue primarily, we suspect, to a low response rate (AAPOR RR3 =

The key questions ask respondents how stronglythey agree with statements about four important val-ues in American political culture: equal opportu-nity, self-reliance, moral traditionalism, and moraltolerance. We selected these values because they representfundamental lines of political cleavage in two of the majorpolicy domains in contemporary American society. First,in the social welfare domain political conflict has longcentered on how best to strike a balance between equalopportunity on one hand and self-reliance on the other(McClosky and Zaller 1984). In a similar vein, politicalconflict in the cultural-moral issues domain has focusedon the tension between the defense of traditional valuesand the acceptance of progressive alternatives (Layman2001). Finally, each of these value conflicts maps onto theparty system in identifiable ways. The Democratic Partyplaces more emphasis on equal opportunity and moraltolerance while the GOP stresses self-reliance and tradi-tional morality.

Each of these principles represents an abstract beliefabout a desirable end-state or mode of conduct. Equal op-portunity is the belief that everyone should have the samechance to succeed and prosper in life. Self-reliance repre-sents the idea that each person should rely on individualinitiative and self-discipline to get ahead. Moral tradition-alism reflects the degree to which one favors customaryor traditional codes of moral behavior and conduct, par-ticularly with respect to family structures. Finally, moraltolerance is the idea that the good and just society ac-cepts individuals and groups whose beliefs and behaviorsregarding right and wrong differ from those of the main-stream (Feldman 1988; Goren 2005). Respondents wereasked to rate how strongly they agree with propositionsregarding each value on a 0–10 scale (higher scores reflectmore support). We employ three statements per value toenhance measurement validity and reliability. Questionwording for each statement appears in Table 1.

Respondents in the control group received the“Some people believe” cue in each value question. Inthe party treatment group, respondents were informedthat “Democrats believe” in equal opportunity andmoral tolerance and “Republicans believe” in self-relianceand moral traditionalism. In the ideology-plus-partytreatment group, respondents were told that “Liberal

.182). Most notably, our sample is more educated than the broaderpopulation. According to Current Population Survey data, roughly25% of the public has a college degree compared to 44% in oursample. While this undermines the representativeness of our sampleto some extent, it makes it harder to demonstrate that partisanbiases shape value positions because the politically sophisticatedare less likely to use partisan cues than the unsophisticated (Kam2005; Lau and Redlawsk 2006).


TABLE 1 Question Wording for the Value Items

Equal OpportunityItem 1: [Some people / Democrats / Liberal Democrats] believe that we need to do a lot more to make sure everyone has

an equal chance to get ahead in life.Item 2: [Some people / Democrats / Liberal Democrats] believe that our country would be much better off if people were

treated more equally.Item 3: [Some people / Democrats / Liberal Democrats] believe that our society should do whatever is necessary to make

sure that everyone has an equal opportunity to succeed.Self-Reliance

Item 1: [Some people / Republicans / Conservative Republicans] believe that individuals should depend more onthemselves and less on others in order to get ahead in life.

Item 2: [Some people / Republicans / Conservative Republicans] believe that our country would be much better off ifthere were more emphasis on self-reliance.

Item 3: [Some people / Republicans / Conservative Republicans] believe that our society would benefit greatly if peoplewere more self-sufficient.

Moral TraditionalismItem 1: [Some people / Republicans / Conservative Republicans] believe that our country would be much better off if

there were more emphasis on traditional family ties.Item 2: [Some people / Republicans / Conservative Republicans] believe that non-traditional lifestyles are contributing to

the breakdown of our society.Item 3: [Some people / Republicans / Conservative Republicans] believe that there should be a lot more respect for

traditional family values.Moral Tolerance

Item 1: [Some people / Democrats / Liberal Democrats] believe that the world is always changing and we should adjustour views of moral behavior to those changes.

Item 2: [Some people / Democrats / Liberal Democrats] believe that we should be more tolerant of people who choose tolive according to their own moral standards.

Item 3: [Some people / Democrats / Liberal Democrats] believe that society should be more accepting of individualswhose values or behaviors are very different from most.

Note: After each statement, respondents were asked to rate how much they agree with it on a 0–10 scale.

Democrats believe” in equal opportunity and moral tol-erance and “Conservative Republicans believe” in self-reliance and traditional values. After hearing these cues,respondents were asked about their own positions oneach value. In sum, control group respondents receivedno cues across the 12 value questions while treatmentgroup respondents encountered six Democratic/liberalDemocratic and six Republican/conservative Republicancues.

Statistical Analyses and SubstantiveResults

Measuring the Value Constructs

We use confirmatory factor analysis (CFA) to estimatethe degree to which the value items are valid and reli-able indicators of our value constructs (all variables have

been normed to 0.00–1.00 scale).6 The model we estimatehas four factors—one for each value—with the items foreach value loading onto the respective factors. For eachfactor, the loading for the first item is set to 1.00. Factorsare coded so higher scores reflect stronger endorsementof the value. We report the standardized factor loadingof each indicator on each factor, Cronbach’s � reliabilitycoefficient for the respective multi-item scales, and stan-dard global fit statistics. The measurement analyses arerestricted to control group respondents since their an-swers to the value items are not contaminated by sourcecues.

The Table 2 estimates reveal that the observed valueitems do a good job tapping latent values. First, all of the

6All latent variable models are estimated in Mplus 4.21. Given thenonnormality of the data, the mean-corrected robust weighted leastsquares (WLSM) estimator is used, with the value items specified asright-censored (Muthen and Muthen 2007). A second estimator fornonnormal data (Satorra-Bentler maximum-likelihood estimationwith robust standard errors) yields similar results.


TABLE 2 Confirmatory Factor Analysis of Value Items

Equal Moral MoralOpportunity Self-Reliance Tradition Tolerance

Standardized Factor Loadings�1 Equal opportunity item 1 .90�2 Equal opportunity item 2 .77�3 Equal opportunity item 3 .76�4 Self-reliance item 1 .70�5 Self-reliance item 2 .80�6 Self-reliance item 3 .94�7 Moral traditionalism item 1 .87�8 Moral traditionalism item 2 .69�9 Moral traditionalism item 3 .88�10 Moral tolerance item 1 .56�11 Moral tolerance item 2 .86�12 Moral tolerance item 3 .77

Cronbach’s � .79 .81 .82 .76

Model FitMean-corrected Robust WLS � 2 160.15Degrees of freedom 47p value <.001CFI .99TLI .99

Notes: N = 294 (control sample). Mean-corrected robust WLS estimation is used, with the values items specified as right-censored. All factor-loading estimates are significant at p < .01. Factor variances, error variances, and error covariancesomitted for clarity.Source: 2006 author survey.

items are valid indicators of their respective constructswith item-factor correlations ranging from a low of .56to a high of .94.7 Next, the individual items are quite re-liable as well (as determined by squaring the item-factorcorrelations): latent egalitarianism explains 58–81% ofthe observed variance in its items, latent self-reliance ac-counts for 49–88% of the observed item variance, latentmoral traditionalism explains 48–77% of the variance inits items, and latent moral tolerance accounts for 31–74%of the observed variance. The respective multi-item scalesare also highly reliable, with Cronbach’s � equaling .79 forequality, .81 for self-reliance, .82 for traditionalism, and.76 for tolerance. Also, while the model � 2 is significant,the other indices reveal excellent global model fit (e.g.,CFI = .99, TLI = .99).

For comparison, we also estimated several alterna-tive models. First, we examined a simple model in whichall 12 value items were taken as indicators of a singlevalue dimension. This alternative model fit the data verypoorly (� 2

53 = 1423.90, p < .0001, CFI = .92, TLI = .86),

7We follow Bollen’s (1989, 199) definition of validity, which equatesit with the correlation between the observed indicator and the latentconstruct it is designed to measure.

with a � 2 difference test indicating that the one-factormodel provided a significantly worse fit than our four-factor model (�� 2

6 = 396.53, p < .0001).8 Second, weexamined the possibility that equality and self-reliancemight be oppositely valenced indicators of a common“economic values” dimension and that moral tradition-alism and tolerance might be oppositely valenced in-dicators of a common “social values” dimension. Thistwo-factor model—which constrained the equality andself-reliance items to load on one factor and the tradi-tionalism and tolerance items to load on the other—also fit poorly (� 2

52 = 1113.31, p < .0001, CFI = .94,TLI = .89), with a � 2 difference test indicating that themodel provided a significantly worse fit compared to ourfour-factor model (�� 2

5 = 258.59, p < .001). Finally, weconsidered an alternative two-factor model that groupedthe values by their left-right flavor. In this model, theequality and tolerance items were allowed to load ona common “left” factor, while the self-reliance and

8The comparison of robust WLS � 2 statistics requires the use of acorrection factor (Muthen and Muthen 2007). As such, the actual�� 2 statistics do not equal the simple difference between the chi-squares for the compared models.


TABLE 3 The Impact of Source Cues on the Partisanship-Value Relationship

Party + IdeologyNo Cue Party Cue Cue

Effect of Partisanship OnEqual opportunity −.49 −.63 −.75Self-reliance .25 .38 .62Moral traditionalism .45 .45 .74Moral tolerance −.24 −.41 −.59

Fit Statistics Unconstrained Multi-group ModelMean-corrected Robust WLS � 2 656.79Degrees of freedom 266p value <.01CFI .99TLI .99

Notes: N = 899 across cells. Variables coded so higher scores denote Republican partisanship and supportfor listed value. All entries are unstandardized path coefficients; all estimates shown are significant at thep < .01 level. Mean-corrected robust WLS estimation is used, with the values items specified as right-censored. Factor loadings, factor variances, error variances, and error covariances omitted for clarity. Allvariables on 0.00–1.00 scale.Source: 2006 author survey.

traditionalism items were allowed to load on a common“right” factor. This model also provided a significantlyworse fit than the four-factor model (� 2

52 = 664.84, p <

.001, CFI = .96, TLI = .94; �� 25 = 151.54, p < .001). In

general, then, the results suggest that a four-factor valuestructure most adequately describes the data, and thateach set of items provides a valid, reliable measure of akey political value.

The Impact of Source Cues on the PartyIdentification-Values Relationship

Our first hypothesis maintains that partisan source cuesactivate Democratic-Republican identifications in theminds of citizens which lead them to take somewhat dif-ferent value positions. If this is to hold empirically, thenexpressed value positions should depend more heavily onpartisan identities in the party and party-plus-ideologytreatment groups relative to the control group. We testthis prediction using structural equation modeling tech-niques in which the four latent values—measured usingthe items as indicated in Table 2—are regressed on latentpartisanship in each of the three experimental groups. La-tent partisanship is measured with three indicators. Thefirst is the standard NES 7-point self-categorization scale.The second and third measures are feeling thermome-ters for the Republican and Democratic parties, with thelatter reverse-coded. These items are scaled from 0 to 10,“with 10 meaning a very warm, favorable feeling, 0 mean-ing a very cold, unfavorable feeling, and 5 meaning not

particularly warm or cold.” Higher scores on latent par-tisanship denote stronger Republican attachments.9 Notefinally that all variables have been normed to run from0.00 to 1.00.

The estimates for the structural model in each con-dition are reported in Table 3. The second column rep-resents the impact of latent partisanship on each valuein the no-cue condition; the third column estimates theimpact of party on each value in the party-cue con-dition (i.e., “Democrats believe” for equal opportunityand moral tolerance and “Republicans believe” for self-reliance and moral traditionalism); the fourth columndoes the same for the party-plus-ideology cues (i.e., “Lib-eral Democrats believe” for equal opportunity and moraltolerance and “Conservative Republicans believe” for self-reliance and traditional values). Let us begin with theparty identification–equal opportunity relationship. Asone would expect, increasing GOP partisanship is nega-tively related to beliefs about equal opportunity in eachgroup. In the control group, movement from the Demo-cratic to the Republican extreme on latent partisanshipis associated with a .49 unit decline in support for equalopportunity. Next, in the party-cue group comparablemovement on latent partisanship leads to a .63 unit de-cline in support. Finally, in the party-plus-ideology groupmovement across latent partisanship is associated with a.75 unit decline. To sum up, the impact of party ties on

9Note that the party identification questions preceded the valuequestions in the survey.


equal opportunity rises monotonically in the presence ofincreasingly explicit party source cues.

A similar pattern of results holds for self-reliance andmoral tolerance and, to a lesser extent, moral traditional-ism. First, Republican partisanship is positively associatedwith support for self-reliance in all three conditions. Thestructural coefficient for latent partisanship equals .25 inthe control group, .38 in the party-cue group, and .62 inthe party-plus-ideology group. Second, Republican tiesare inversely related to moral tolerance, with the effectrising in magnitude from −.24 to −.41 to −.59 acrossthe respective conditions. Third, moral traditionalism ispositively related to Republican partisanship. Its depen-dence on partisanship is the same in the control andparty-cue groups; however, when respondents hear thatconservative Republicans endorse family values, the im-pact of partisan loyalties on value positions grows from.45 to .74.

Broadly speaking, the Table 3 estimates suggest thatpartisan source cues lead individuals to rely more heav-ily on partisan identities to vertically constrain their corepolitical principles. Likewise, the differences in the mag-nitudes of the party coefficients across the conditions arein the expected direction and substantively significant.To formally test whether the party identification coeffi-cients differ across the groups, we estimate two additionalmodels in which the coefficients for the effect of partyidentification are constrained to equality for (1) no-cueand party-cue respondents and (2) no-cue and party-plus-ideology cue respondents. Robust WLS � 2 differencetests indicate that the constraints produced a significantdecrease in fit for the no-cue versus party comparison(overall fit: � 2

270 = 654.04, p < .01, CFI = .99, TLI = .99;�� 2

4 = 9.17, p = .057) and for the no-cue versus party-plus-ideology comparison (overall fit: � 2

270 = 866.72,p < .01, CFI = .99, TLI = .99; �� 2

4 = 31.48, p < .001).In conjunction, these estimates reveal that partisan cuesstrengthen vertical connections between partisan loyaltiesand beliefs about equal opportunity, self-reliance, moraltolerance, and, to a lesser extent, moral traditionalism,thereby supporting our first hypothesis.10

In these analyses, we rely on omnibus tests since weare interested in across-group differences in the overall re-lationship between partisanship and the group of values asa whole. However, in order to see what relationships were

10As a final comparison, we estimated a model constraining theparty identification coefficients in the control group to equal theaverage of the corresponding party identification coefficients inthe two treatment groups. This model examines whether the coef-ficients in the cued treatment conditions differ from those in thecontrol, pooling across conditions. Robust WLS � 2 difference testsindicate that this constraint also significantly decreased the fit ofthe model (�� 2

4 = 28.16, p < .001), lending further support to ourfirst hypothesis.

most implicated in these across-group differences, we alsoconducted analyses that constrained the party-value co-efficients to equality across groups one at a time. Forthe control versus party comparison, these tests indicatedthe omnibus difference was driven primarily by groupdifferences in the effects of partisanship on self-reliance(p < .10) and tolerance (p < .05). For the control ver-sus party-plus-ideology comparison, these tests indicatethe omnibus difference was driven by group differencesin the effects of partisanship on all four values, but es-pecially self-reliance and tolerance (p < .06 for equality;p < .001 for self-reliance; p < .06 for traditional values;p < .001 for tolerance).

In-Party versus Out-Party Cues

The next step is to test the negativity bias prediction thatout-party cues will produce bigger net changes in valueorientations than in-party cues. We cannot use structuralequation modeling techniques here because the samplesizes in some cells are too small (≈25–50) to generatestable parameter estimates. Instead, a series of dummyvariable regression models are specified whereby the inde-pendent variables reflect different configurations of partyidentification and experimental conditions. In estimatingthese models, we also account for the possibility that thehigh level of endorsement for some values may have cre-ated a situation where responses were “right-censored”or subject to a ceiling effect. Many respondents endorsedeach value to the maximum possible extent. This “bunch-ing” at the top end of the scale suggests that the scale itselfmay have imposed an artificial ceiling on responses andthat respondents at the maximum observed point mayactually differ in terms of latent support for the value,with some endorsing it more than others. In cases likethis, the use of ordinary least squares can produce incon-sistent parameter estimates biased toward zero (Greene2003; Long 1997). Therefore, we estimate our models us-ing right-censored maximum-likelihood tobit regression,which is appropriate for dependent variables where a ceil-ing effect is suspected (Greene 2003). This approach “fillsin” the missing information about variance in the depen-dent variable at its ceiling by incorporating informationabout different cases’ probability of being censored at thetop of the scale, thereby producing consistent estimatesof the effects of the independent variables (Long 1997).11

11Right-censored tobit regression allows predicted values to exceedthe upper bound of the original dependent variable (i.e., y > 1.00).These estimates indicate what the predicted value would be if theresponses were not censored at the upper bound. Some of ourTables 4–5 results confirm that ceiling effects are present, as thetobit estimates generate predicted values in excess of 1.00.


The critical question is whether Republicans re-spond more forcefully than Democrats when cued that“Democrats believe” in equal opportunity and moral tol-erance, and whether Democrats respond more vigorouslythan GOP partisans when hearing that “Republicans be-lieve” in self-reliance and moral traditionalism.12 If neg-ativity bias is present, then the absolute value of the de-crease in expressed value support for partisans gettingthe out-party cue should exceed the increase among thecue giver’s fellow partisans.13 We explore these patternsfor partisans in general (i.e., strong, weak, and leaningidentifiers on the 7-point self-categorization scale) andextreme partisans (i.e., strong identifiers on the 7-pointscale).

Table 4 presents the predicted mean level of expressedsupport for each value (on a 0.00–1.00 scale) among iden-tifiers in the control and party-cue groups. We begin withequal opportunity. For Democrats the equal opportunitymean is .89 in the uncued group and .93 in the cued group,indicating that the Democratic source cue enhances ex-pressed support among Democratic partisans (p < .10).As predicted, the source cue proves more consequentialamong Republicans as their expressed support for equal-ity declines from .73 in the control group to .63 in thetreatment group (p < .05).14 The same pattern holdsfor strong partisans. When strong Democrats hear thatDemocrats believe in equal opportunity, their expressedsupport rises from .96 in the uncued group to 1.02 inthe cued group (p < .10). When strong Republicans re-ceive what is for them a negatively valenced cue, expressedsupport for equal opportunity falls from .70 to .60 (p <

.05). In short, the Democratic source cue produces a big-ger net change in value endorsement among Republicansthan among Democrats.

12Recall that negativity bias will be present when the perceptualbackground is evaluated positively and is accompanied by a neg-atively valenced cue. We assume that in-party/out-party cues arepositively/negatively valenced for those partisans. Moreover, ouranalysis shows that the mean level of support for each value ispositively valenced (i.e., value mean ≥ .50) for Democratic andRepublican identifiers (leaners counted as partisans) in the controlcondition in 13 of the 16 group comparisons we make in Tables 4–5.The exceptions are strong Republicans, conservative Republicans,and strong conservative Republicans on moral tolerance.

13Our experimental design only allows us to test whether differ-ences in cue valence across partisan categories influence expressedvalue support (e.g., the impact a Democratic endorsement has onDemocratic and Republican value positions). We cannot explorewhether differences in cue valence within partisan categories pro-duces asymmetrical responses (e.g., the impact Democratic andRepublican sanctions have on Democratic value positions). Onceagain, ours is not a complete test of the negativity bias hypothesis.

14All results remain very similar in form, magnitude, and signifi-cance when controls for age, income, gender, education, and raceare included.

Does the same pattern obtain when Republicansources endorse self-reliance? When Democrats receivethis message, expressed support declines from .72 inthe control group to .63 in the treatment group (p <

.01). The decline is even more pronounced among strongDemocrats as the Republican cue yields a .77 to .61 de-cline (p < .01). In contrast, the Republican endorsementdoes not systematically affect positions taken by Republi-can identifiers. The cue leads to a decline in expressedsupport among Republicans (p < .05); however, thiscounterintuitive effect vanishes among strong partisans(.91 vs. .90, ns). More broadly, the self-reliance experi-ment demonstrates that out-party cues have more po-tent effects on identifiers’ expressed values than in-partycues.

The results for the moral traditionalism experimentsappear in the third panel of Table 4. Here we find morelimited evidence that party cues motivate changes in ex-pressed value support among opposition partisans. Toelaborate, Republican endorsement of family values hasno effect in the larger partisan groups. Cued Democratsare no less willing to endorse family values than uncuedDemocrats (.61 vs. .62, ns); the same holds true for GOPidentifiers (.86 vs. .84, ns). When attention turns to strongpartisans, Republican endorsement of traditional moral-ity drives expressed Democratic support down from .68 to.59 (p < .10) and has no effect on the views of ardent Re-publicans (.98 vs. .96, ns). In short, negativity bias seemsto emerge only among strong partisans.

Moral tolerance is the final value under consider-ation. These experiments do not support our expecta-tions. First, the Democratic source cue does not appear toinfluence expressed value support among Democrats orRepublicans. Second, among strong identifiers the sourcecue effect is again limited to one side of the spectrum,in this case increasing expressed support among those onthe left. When strong Democrats are told their copartisansbelieve in tolerating those with diverse moral viewpoints,tolerance scores rise from .68 to .78 (p < .05). StrongRepublicans, in contrast, prove unresponsive, with thosein the cued group as cool toward tolerance as those in theuncued group (.45 vs. .43, ns).

To summarize the results so far, out-party cues pro-duce larger effects on expressed value support among par-tisans in general on two of four values (equal opportunityand self-reliance) and among strong partisans on three offour values (equal opportunity, self-reliance, and moraltraditionalism). So far, then, the negativity bias hypothe-sis receives real if not overwhelming support. However, aswe shall see shortly, evidence on its behalf is even strongerwhen analysis is confined to ideologically congruentpartisans.


TABLE 4 Predicted Support for Political Values among Partisans

No Cue Party Cue Cue-No Cue

Equal OpportunityDemocrats .89 .93 .04+

Republicans .73 .63 −.10∗

Rep – Dem −.16∗∗ −.30∗∗

−2LL = 269.06, � 2(3) = 123.79, p < .001; n = 577Strong Democrats .96 1.02 .06+

Strong Republicans .70 .60 −.10∗

Rep – Dem −.26∗∗ −.42∗∗

−2LL = 56.58, � 2(3) = 106.42, p < .001; n = 206Self-Reliance

Democrats .72 .63 −.09∗∗

Republicans .90 .85 −.05∗

Rep – Dem .18∗∗ .22∗∗

−2LL = 117.76, � 2(3) = 118.98, p < .001; n = 581Strong Democrats .77 .61 −.16∗∗

Strong Republicans .91 .90 −.01Rep – Dem .14∗∗ .29∗∗

−2LL = 86.84, � 2(3) = 50.74, p < .001; n = 213Moral Traditionalism

Democrats .62 .61 −.01Republicans .84 .86 .02

Rep – Dem .22∗∗ .25∗∗

−2LL = 336.62, � 2(3) = 92.56, p < .001; n = 581Strong Democrats .68 .59 −.09+

Strong Republicans .98 .96 −.02Rep – Dem .30∗∗ .37∗∗

−2LL = 168.46, � 2(3) = 56.84, p < .001; n = 212Moral Tolerance

Democrats .67 .70 .03Republicans .51 .48 −.03

Rep – Dem −.16∗∗ −.22∗∗

−2LL = 103.76, � 2(3) = 80.47, p < .001; n = 571Strong Democrats .68 .78 .10∗

Strong Republicans .43 .45 .02Rep – Dem −.25∗∗ −.33∗∗

−2LL = 59.56, � 2(3) = 64.67, p < .001; n = 210

+ p < .10; ∗p < .05; ∗∗p < .01.Notes: Democratic cue present for equal opportunity and moral tolerance; Republican cue present for self-reliance andmoral traditionalism. Cell entries represent predicted means generated from dummy variable tobit-regression equations.All variables on 0.00–1.00 scale.Source: 2006 author survey.

The Role of Ideological Congruence

The next step in the analysis is to test the hypothesis thatthe interaction between source cues and partisan loyaltiesis stronger for ideologically pure partisans than for parti-san groups containing those of all ideological stripes. We

test this hypothesis by estimating tobit regression modelsfor self-identified liberal Democrats and conservative Re-publicans in the control group (i.e., no cue) and secondtreatment group (i.e., the “Liberal Democrats” or “Con-servative Republicans” cues) and comparing these resultswith those reported for the ideologically mixed samplesin Table 4.


Liberal-conservative identification is measured us-ing a 7-point self-categorization scale.15 We use responsesto both the party and ideology self-identification scalesto classify individuals into the following comparativegroups: (1) liberal Democrats versus conservative Re-publicans and (2) strong liberal Democrats versus strongconservative Republicans. The first pairing distinguishesanyone on the left or right side of both divides; the seconddistinguishes those at the left or right extremes.

Table 5 contains the value means for all relevantgroups. Overall, the results are by and large consistentwith expectations. First, changes in expressed value sup-port are typically larger for ideologically congruent par-tisans compared to ideologically mixed partisans. To takeone example, in the presence of a left-wing cue, supportfor equal opportunity falls more among the ideologicallypure strong Republicans in Table 5 (from .67 to .52, p <

.01) compared to the ideologically mixed strong Republi-cans in Table 4 (from .70 to .60, p < .05). To take anotherexample, when the GOP endorses traditional morality,support from strong liberal Democrats in Table 5 fallsmore (.50 to .33, p < .01) than that of the ideologicallymixed Democrats in Table 4 (.68 to .59, p< .10). The pointis clear. Generally speaking, source cues produce moresubstantial changes in expressed value support when cit-izens’ liberal-conservative leanings reinforce their partyloyalties. These results lend solid support to our fourthhypothesis.

The second point to emphasize is that out-party cuesagain prove more powerful than in-party cues. To illus-trate, the “liberal Democrats” cue for equal opportunityleads to big declines in expressed support for this valueamong conservative Republicans (from .71 to .58, p <

.01) and strong conservative Republicans (from .67 to

.52, p < .01). In contrast, liberal Democrats and strongliberal Democrats hold steady in their expressed valuesupport (.90 to .92, ns, and 1.05 to 1.05, ns, respectively).To take another case, the conservative Republican cue forself-reliance significantly reduces expressed liberal Demo-cratic support (.69 to .60, p < .01) and does not influencethe support offered by conservative Republicans (.89 vs.91, ns). Furthermore, the cue produces a much larger dropin expressed support among strong liberal Democrats (.71to .53, p < .01) versus the smaller rise in expressed sup-port among strong conservative Republicans (.93 to 1.02,

15Respondents were asked, “Generally speaking, do you usuallythink of yourself as a liberal, a conservative, a moderate, or haven’tyou thought much about this?” Those selecting “liberal” or “conser-vative” received a follow-up query: “Would you call yourself a strong[liberal/conservative] or a not very strong [liberal/conservative]?”Everyone else is asked, “Do you think of yourself as closer to liberalsor closer to conservatives?”

p < .10). When we take a step back to consider the big-ger picture, the tobit estimates reveal that out-party cuesproduce bigger shifts on three of four values for both ideo-logically pure partisans (equal opportunity, self-reliance,and moral tolerance) and strong ideologically pure par-tisans (equal opportunity, self-reliance, and moral tra-ditionalism).16 In conjunction, the results from Tables 4and 5 support the negativity bias hypothesis in 11 of 16tests.17

Value Constraint

Our final hypothesis predicts that levels of constraintamong political values will rise in the presence of partysource cues. If we can demonstrate that party cuesstrengthen our ability to predict support for one valuefrom another, then we can be more confident that thechanges produced by the cues have important effects be-yond shaping expressed value positions. To test this pre-diction, we estimate a multigroup version of the four-factor model shown in Table 2 for the control group, theparty group, and the party-ideology group. The key es-timates we need are the correlations between the latentfactors in each experimental group. These estimates arereported in Table 6.

A quick glance at the factor correlations revealsthat core political values are indeed more stronglyinterrelated—in the predicted directions—in the twotreatment groups. While the mean of the absolute value ofthe correlations in the no-cue condition is .27, the equiv-alent averages in the party-cue and party-plus-ideologycue conditions are .45 and .41, respectively. These resultssuggest that most of the increase in value constraint is

16In Tables 4–5, one might ask whether each observed differencebetween the Cue-No Cue means for Democrats and Republicans issignificant (e.g., in Table 4, equality opportunity panel, is the .04difference across conditions for Democrats significantly differentfrom the −.10 difference across conditions for Republicans). Wefind that the preponderance of the evidence continues to supportour claims. To elaborate, additional analysis (available from the au-thors upon request) revealed that of the 13 differences highlightedin Tables 4–5, eight of these are statistically significant and twomore approach significance. Note that in some cases where signif-icance levels were marginal or insignificant, sample sizes are quitesmall (i.e., n ≈ 100–200). Hence, the results continue to supportour claims in 10 of the 13 tests. Moreover, the supplemental analy-ses continue to show that larger cue effects emerge in the party +ideology cue compared to the party-cue condition.

17Recall from note 12 that the moral tolerance control group meanis less than .50 for strong Republicans, conservative Republicans,and strong conservative Republicans. Since negativity bias shouldoperate when the perceptual background is positive, a case can bemade that we should not expect negativity biases to be present inthese three instances. If we adopt this position, our results yieldevidence of negativity bias in 10 of 13 tests.


TABLE 5 Predicted Support for Political Values among Ideologically CongruentPartisans

No Cue PID + Ideo. Cue Cue-No Cue

Equal OpportunityLiberal Democrats .90 .92 .02Conservative Republicans .71 .58 −.13∗∗

Rep – Dem −.19∗∗ −.34∗∗

−2LL = 126.72, � 2(3) = 109.49, p < .001; n = 372Strong Liberal Democrats 1.05 1.05 .00Strong Conservative Republicans .67 .52 −.15∗∗

Rep – Dem −.38∗∗ −.53∗∗

−2LL = 23.40, � 2(3) = 69.91, p < .001; n = 99Self-Reliance

Liberal Democrats .69 .60 −.09∗∗

Conservative Republicans .89 .91 .02Rep – Dem .20∗∗ .31∗∗

−2LL = 90.06, � 2(3) = 111.28, p < .001; n = 379Strong Liberal Democrats .71 .53 −.18∗∗

Strong Conservative Republicans .93 1.02 .09+

Rep – Dem .22∗∗ .49∗∗

−2LL = 52.26, � 2(3) = 45.92, p < .001; n = 105Moral Traditionalism

Liberal Democrats .52 .50 −.02Conservative Republicans .87 .89 .02

Rep – Dem .35∗∗ .39∗∗

−2LL = 126.72, � 2(3) = 148.81, p < .001; n = 376Strong Liberal Democrats .50 .33 −.17∗

Strong Conservative Republicans 1.00 .96 −.04Rep – Dem .50∗∗ .63∗∗

−2LL = 55.50, � 2(3) = 74.40, p < .001; n = 103Moral Tolerance

Liberal Democrats .73 .74 .01Conservative Republicans .48 .39 −.09∗∗

Rep – Dem −.25∗∗ −.35∗∗

−2LL = 126.72, � 2(3) = 154.01, p < .001; n = 371Strong Liberal Democrats .72 .86 .14+

Strong Conservative Republicans .39 .33 −.06Rep – Dem −.33∗∗ −.53∗

−2LL = 8.00, � 2(3) = 59.98, p < .001; n = 102

+ p < .10; ∗ p < .05; ∗∗ p < .01.Notes: Liberal Democratic cue present for equal opportunity and moral tolerance; Conservative Republican cue presentfor self-reliance and moral traditionalism. Cell entries represent predicted means generated from dummy variable tobit-regression equations. All variables on 0.00–1.00 scale.Source: 2006 author survey.

driven by the party cue. To formally test whether the fac-tor correlations differ in magnitude across the groups, weestimate two additional models in which the correlationsare constrained to equality for (1) no-cue and party-cuerespondents and (2) no-cue and party-plus-ideology cuerespondents. Robust WLS � 2 difference tests indicate that

the constraints produced highly significant decreases infit for the no-cue versus party-cue comparison (over-all fit: � 2

163 = 619.09, p < .001, CFI = .99, TLI = .99;�� 2

6 = 29.63, p = .001) and the no cue versus party-plus-ideology comparison (overall fit: � 2

163 = 723.98,p < .001, CFI = .99, TLI = .99; �� 2

6 = 41.81, p < .001).


TABLE 6 Group Differences in Factor Correlations Between Values

Equal Moral MoralOpportunity Self-Reliance Tradition Tolerance

No CueEqual Opportunity 1.00Self-Reliance −.02 1.00Moral Traditionalism −.13+ .50∗∗∗ 1.00Moral Tolerance .46∗∗∗ −.09 −.42∗∗∗ 1.00

Party CueEqual Opportunity 1.00Self-Reliance −.29∗∗∗ 1.00Moral Traditionalism −.50∗∗∗ .66∗∗∗ 1.00Moral Tolerance .62∗∗∗ −.14∗ −.50∗∗∗ 1.00

Party + Ideology CueEqual Opportunity 1.00Self-Reliance −.34∗∗∗ 1.00Moral Traditionalism −.15∗∗ .59∗∗∗ 1.00Moral Tolerance .59∗∗∗ −.29∗∗∗ −.50∗∗∗ 1.00

Fit Statistics Unconstrained Multi-group ModelMean-corrected Robust WLS � 2 517.47Degrees of freedom 157p value <.001CFI .99TLI .99

+p < .10; ∗p < .05; ∗∗p < .01; ∗∗∗p < .001.Notes: All entries are standardized factor correlations. Mean-corrected robust WLS estimation is used, withthe values items specified as right-censored.Source: 2006 author survey.

Hence, the cues provided to our respondents not onlyinduce changes in their expressed value positions, butalso induce changes in the deeper structure of their val-ues. In the presence of cues about “who” supports dif-ferent values, mass belief systems become more tightlyconstrained.

Conclusions, Qualifications,and Implications

For nearly half a century research on party identifica-tion has trod along familiar grounds. Most studies esti-mate statistical models using nonexperimental NES datato gauge how partisan loyalties shape retrospective judg-ments, candidate evaluations, and other short-term per-ceptions (e.g., Bartels 2002; Goren 2002). A few worksutilize experimental designs to isolate the causal impactparty ties have on short-term perceptions (Rahn 1993;Lau and Redlawsk 2006). To the best of our knowledgeno prior work employs experimental designs to assess

the causal impact party identification has on long-termpredispositions.

We have argued that partisan source cues in general—and opposition-party cues in particular—evoke partisanbiases in the minds of citizens that subsequently influenceexpressed support for equal opportunity, self-reliance,moral traditionalism, and moral tolerance. We furtherposited that the source cue–party identity effects will bemore pronounced when ideological sentiments reinforcepartisan leanings. Finally, we expected party cues to pro-mote horizontal constraint among these values. Thesepropositions were tested via a series of novel survey ex-periments administered to an adult cross-section of theU.S. population.

Four compelling findings have emerged from theanalyses. First, the relationship between individual par-tisan identities and expressed value support is strongerin the presence of party source cues. Second, out-partycues typically prove more powerful than in-party cues inshaping expressed value support. Third, these relation-ships are usually more pronounced among ideologically


congruent partisans than among partisans of all ideolog-ical stripes. Fourth, party cues promote horizontal con-straint among these values. The source cues did not alwaysinteract with political identities in the manner predictedand some of the reported effects are of modest magnitude;nevertheless, the general pattern of results suggests thatpartisanship has the power to shape expressed supportfor—and the cognitive organization of—political valuesunder a variety of conditions.

As is the case for most survey experiments, ques-tions can be raised about the generalizability of our re-sults beyond the research setting. Our work shows thatthe presence of party cues leads partisans to alter theirresponses to value questions in the context of a singlesurvey. It does not demonstrate that party cues lead tofundamental changes in latent value positions that en-dure over the long haul. Repeated exposure to consistentpartisan messages on party-value linkages would be nec-essary to produce lasting value change. Of course, if thelatter type of change is to occur at all, we must observeprecisely the types of changes in value expression we havefound here.

We believe the results of our survey experiments havereal-world analogues for the following reasons. If parti-san political elites take consistent positions on politicalprinciples for an extended period of time, it seems fairto posit that the movements we uncovered in our ex-periments will occur in the minds of citizens outside thesurvey context. Although our work cannot speak directlyto this, some evidence corroborates our claim. With re-spect to political elites, work on polarization indicatesthat the national parties have taken increasingly distinctideological profiles in recent years. Democratic elites havemoved to the left, Republicans to the right, on most is-sues (Hetherington 2001; McCarty, Poole, and Rosenthal2006). Presumably, this holds for the types of values dis-cussed here. Insofar as elite cues compel ordinary citizensto think in partisan terms, one might expect that citizensuse their partisan ties to update their value positions.As mentioned above, this is precisely what Goren (2005)finds. He shows that party identification leads to valuechanges two-to-four years out. Although Goren does nottest for the underlying mechanism driving these results,the claim that elite influence is paramount seems highlyplausible. In sum, since our experimental findings areconsistent with those generated by nonexperimental re-search situated over much longer time frames, we havesome confidence that our results represent more thanexperimental artifacts.

Another objection to our claim is that party cues in-duce nothing more than a temporary change in valueexpression based on how people interpret value mes-sages. To elaborate, political values may be vulnerable

to the influence of source cues because they lack the cog-nitive support necessary to inoculate them from attack(Maio and Olson 1998a, 1998b). Since values are self-evidently good, there is no need to develop reasons forsupporting them. This renders political values suscepti-ble to short-term partisan influence because citizens canuse the source cues to fill in the missing information.18

To illustrate, a declaration that “some people” supportequal opportunity may imply something different from aclaim that “some Democrats” support equal opportunity.The first statement represents a fundamental componentof the American political creed. In the second statement,respondents may interpret the party cue as implying sup-port for helping African Americans and then use thisinformation to guide their value expressions. The sourcecue, in essence, supplies the missing cognitive support (cf.Kuklinski and Hurley 1994). However, once the persua-sive attempt has passed, the value expression is forgottenand the latent value resets to its default position.

Although this represents a plausible account as towhat occurs in a single instance, we do not believe thatit undermines our fundamental conclusion. In isolation,one cue is unlikely to produce lasting change. In con-junction, a series of reinforcing cues should chip awayat the initial value position precisely because this posi-tion lacks much in the way of cognitive justification. Thekey to the process, then, would be for the parties andtheir spokespersons to take consistent left-right positionson political values. The polarization research cited abovesuggests that they have done so in recent decades. Again,even if citizens’ interpretations of party-cued value mes-sages lead to short-term changes in political value expres-sion rather than in latent value positions, the cumulativeimpact of a series of party cues over time should lead tolong-term modification of core political principles.

Next, we reiterate that partisanship does not moti-vate citizens to abandon their political values. Our find-ings suggest partisanship influences value support at themargins rather than inducing wholesale change. Republi-cans do not reject equal opportunity when informed thatDemocrats believe in it, nor do Democrats abandon self-reliance when told that Republicans are for it; instead,expressed support for these values cools slightly. Hence,no claim is made that political values are nonessential el-ements in mass belief systems, on par with judgmentsabout presidential performance or evaluations of theeconomy. Our claim is simply that the political valuepositions citizens are willing to endorse depend partly onDemocratic-Republican identifications.

18We thank an anonymous reviewer for bringing this point to ourattention.


We conclude by noting some of the broader impli-cations our findings have for the study of public opinionand political psychology. First, the findings complementnonexperimental studies documenting partisan bias overa wide range of other political attitudes, beliefs, and be-haviors (Bartels 2002; Green and Palmquist 1990). Ourcontribution on this score is showing that party affectsnormative political beliefs. Second, the evidence that out-party cues exert more influence over value judgments thanin-party cues complements a large body of research doc-umenting negativity bias in politics across diverse areas,such as evaluations of political candidates (e.g., Goren2002), evaluations of political institutions (e.g., Grosskopfand Mondak 1997), and retrospective voting (Hethering-ton 1996), to name but a few. It seems clear that “bad”party cues are stronger than “good” party cues (Baumeis-ter et al. 2001). Finally, our results speak to questionsregarding the nature of human belief. While it is true thatpeople can work out what they believe is best for soci-ety irrespective of their group loyalties, it seems fair tosay that group attachments usually weigh rather heavilyon such beliefs (Greene 1999). In the domain of publicaffairs, a citizen’s commitment to a particular vision ofthe good and just society may be swayed by what partisanfriends and foes believe.

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Source Cues, Partisan Identities, and Political Value ...users.polisci.umn.edu/~pgoren/Goren AJPS 09.pdf · Source Cues, Partisan Identities, and Political Value Expression Paul Goren