What is political efficacy?

The concept “sense of political efficacy” was developed by Campbell et al. (1954, p. 187), who defined it as “the feeling that individual political action does have, or can have, an impact upon the political process, i.e., that it is worthwhile to perform one’s civic duties. It is the feeling that the individual citizen can play a part in bringing about this change.”

The focus of Campbell and colleagues’ research was on perceived political efficacy as a potential explanatory variable for variation in electoral participation. Later work on the construct diverged into two different perspectives: One line of research focused on the importance of political efficacy for the individual (e.g., Renshon, 1975); the other focused on the importance of political efficacy for the political system (e.g., Almond & Verba, 1963).

Internal and external political efficacy

Campbell et al. (1954) conceived “sense of political efficacy” as a unidimensional construct. Balch (1974) was the first to empirically examine the factor structure of a slightly modified version of Campbell et al.’s original measure of the concept. He extracted two components of “sense of political efficacy”: internal political efficacy (personal effectiveness; Craig & Maggiotto, 1982) and external political efficacy (system responsiveness; Craig & Maggiotto, 1982). More specifically, internal political efficacy refers to the belief that one is capable of understanding politics and engaging in political acts. In contrast, external political efficacy refers to the individual’s belief in the responsiveness of political institutions and leaders to citizens’ concerns (W. E. Miller & Traugott, 1989).

The relationship between internal and external efficacy is typically weak because these two components of political efficacy have different referents: oneself (in the case of internal political efficacy) and politicians and political institutions (in the case of external political efficacy; Craig & Maggiotto, 1982; W. E. Miller & Traugott, 1989). Previous findings have shown that internal and external political efficacy are positively associated with each other in democratic contexts (e.g., Craig, 1979; Morrell, 2003; Muller, 1970; Niemi et al., 1991) but largely independent of each other in authoritarian regimes (e.g., Muller, 1970). For instance, Muller (1970) found a medium correlation between internal and external political efficacy in the United States but an almost zero correlation in Mexico. This suggests that, through socialization, people in democratic regimes learn to believe in the responsiveness of their governments (see also Easton & Dennis, 1967). The existence of a norm of regime responsiveness encourages people to develop the skills and, accordingly, the self-conviction to be able to exert influence on governmental decisions (Muller, 1970). In contrast, people in authoritarian regimes may experience a discontinuity in regime responsiveness during their socialization process, which hinders them from acquiring a stable political norm of regime responsiveness. Although people in authoritarian regimes may possess the skills and, accordingly, the conviction to influence politics, the skills and conviction do not match reality. Due to the authoritarian nature of the regime, people are unable to make use of their political skills. They cannot exert political influence; the government does not respond to their concerns (Muller, 1970). Thus, the strength of the association between internal and external political efficacy varies with regime type.

Internal and external political efficacy also show different patterns of associations with external correlates (e.g., potential causes or effects). On the one hand, internal political efficacy is robustly associated with participation in campaign- and community-oriented activities (e.g., Craig, 1979; Morrell, 2003; Niemi et al., 1991). High internal political efficacy corresponds also with (positive attitudes toward) political protest (e.g., Balch, 1974; Craig & Maggiotto, 1982; Šerek et al., 2017). Thus, participatory experiences seem to boost individuals’ perceived political effectiveness (Šerek et al., 2017). Furthermore, high educational attainment, social status, personal interest, political attentiveness, and political information lead to stronger convictions of internal political efficacy (e.g., Craig & Maggiotto, 1982; Morrell, 2003; Niemi et al., 1991). Internal political efficacy is also related to personality: People who are high in Openness are commonly critical thinkers; people who are high in Extraversion are often dominant, assertive, and persuasive. Thus, these two Big Five traits are commonly associated with (internal) political efficacy (Vecchione & Caprara, 2009).

Convictions about external political efficacy have been found to vary as a function of political context and regime performance (Coleman & Davis, 1976). For example, a nation with a functioning democratic system elicits a strong perception of external political efficacy (Coleman & Davis, 1976). External political efficacy is also closely connected to political trust (e.g., Balch, 1974; Craig, 1979; Niemi et al., 1991). A. H. Miller (1974) defined political distrust as the belief in a dysfunction of the political system in which political outputs diverge from the public’s interest. Thus, it is reasonable that people put trust in their government as long as they are convinced that it responds to the public’s concerns—that is, as long as people have a strong conviction of external political efficacy (Craig & Maggiotto, 1982). In contrast, high willingness to engage in political protest is associated with low external political efficacy (Balch, 1974; Šerek et al., 2017), which may be due to “disappointing outcomes, in which the demands of the protesting citizens were not fulfilled” (Šerek et al., 2017, p. 354).

In this study, we evaluated PESS, the English-language adaptation of PEKS, in the UK, and investigated its comparability with the German-language source version in Germany. We analyzed descriptive statistics and psychometric quality criteria—more precisely, objectivity, reliability, and validity—in both language versions. Moreover, we assessed test fairness across both nations with measurement invariance tests.

Objectivity

Researchers should be able to apply, evaluate, and interpret PESS/PEKS objectively. As PESS/PEKS contains fixed written instructions, a standardized order of the items, and a fixed number of labeled categories, it can be applied objectively. PESS/PEKS data can also be evaluated objectively because the scale is accompanied by strict rules on how to model it and how to build sum scores. PESS/PEKS can be interpreted objectively because reference values (i.e., descriptive statistics) have been provided.

Reliability

To estimate the reliability of the two PESS/PEKS subscales (internal and external political efficacy), we computed McDonald’s omega (McDonald, 1999; Raykov, 1997) and test–retest stability. McDonald’s omega was relatively high for both the UK and Germany (UK: ω_Internal = .88, ω_External = .84; DE: ω_Internal = .86, ω_External = .86). The test–retest stability (UK: r_{tt – Internal} = .83, r_{tt – External} = .68; DE: r_{tt – Internal} = .87, r_{tt – External} = .75) after approximately 3 to 4 weeks (Mdn_UK = 28 days; Mdn_DE = 20 days) was generally lower than McDonald’s omega in both nations, but it was still acceptable.

In sum, PESS/PEKS achieved internal consistencies and test–retest stabilities ranging from .68 to .88 in both nations. Especially in view of the fact that PESS/PEKS comprises only two items per subscale, these reliabilities are relatively high and are sufficient for research purposes (e.g., Aiken & Groth-Marnat, 2006; Kemper et al., 2019).

Validity

We investigated the nomological network of PESS/PEKS. As the nomological network was computed based on manifest correlations, the reported values represent lower-bound estimates of the true associations. The correlation coefficients are depicted in Table 5; their interpretation is based on Cohen (1992), who differentiated between a small (r ≥ .10), medium (r ≥ .30), and strong effect (r ≥ .50). Due to alpha accumulation through multiple testing, only coefficients with a significance level of p < .001 are interpreted (Table 5 displays unadjusted p values). In order to investigate the nomological network, we correlated PESS/PEKS with the constructs outlined in the Material section above.

As outlined in the Theoretical background section, internal political efficacy should be associated with personality (especially Big Five Openness and Extraversion; e.g., Vecchione & Caprara, 2009). In line with this expectation, we found that, in both nations, respondents who scored high in Openness and/or Extraversion (see also Beierlein, Kemper, et al., 2014b; Cooper et al., 2013; Vecchione & Caprara, 2009), and respondents who scored high in Emotional Stability (see also Beierlein, Kemper, et al., 2014b), had higher degrees of internal political efficacy. Conscientiousness was negatively related to internal political efficacy in Germany. External political efficacy was uncorrelated to (almost) all Big Five dimensions (see also Beierlein, Kemper, et al., 2014b).

We also assumed that political efficacy would be positively associated with general self-efficacy because it is a specific manifestation of this trait (Condon & Holleque, 2013). Our findings corroborated that expectation with respect to internal political efficacy (see also Beierlein, Kemper, et al., 2014b; Condon & Holleque, 2013). However, external political efficacy was unrelated to general self-efficacy (see also Beierlein, Kemper, et al., 2014b).

High political efficacy suggests a “feeling of mastery over political processes” (Shrivastava, 1989, p. 171; see also Minton, 1972). Thus, high internal political efficacy should be associated with a high internal locus of control. This expectation was corroborated by our data (see also Beierlein, Kemper, et al., 2014b; Sigel, 1975). Other associations differed across the two nations: Whereas internal political efficacy was positively related to external locus of control and external political efficacy was positively related to both internal and external locus of control in the UK, we did not find any associations between political efficacy and locus of control in Germany.

One might also suspect that optimism as a personality trait pervades people’s sense of (political) efficacy (e.g., Campbell et al., 1954, 1960; Shrivastava, 1989). In line with this reasoning, we found that, in both the UK and Germany, respondents with high internal political efficacy were more optimistic than those with low internal political efficacy. External political efficacy had no significant association with optimism.

Table 5

Correlations of Internal and External Political Efficacy with Relevant Variables in the UK and German Samples

Note. UK = United Kingdom (N = 468; N_{Left–right self-placement} = 325; N_Employment = 339; N_Income = 431); DE = Germany (N = 474; N_{Left–right self-placement} = 394; N_Employment = 309; N_Income = 449); CI = confidence interval; PQ+ = Exaggerating positive qualities; NQ‒ = Minimizing negative qualities. NQ‒ was recoded so that high scores on PQ+ and high scores on NQ‒ imply higher socially desirable responding. Optimism–pessimism: very pessimistic (1) – very optimistic (7). Employment: 1 = unemployed, 2 = employed. Gender: 1 = male, 2 = female. ^*p < .05, ^**p < .01, ^***p < .001.

As outlined in the Theoretical background section, external political efficacy should correlate positively with trust (e.g., Balch, 1974; Craig, 1979; Niemi et al., 1991). We corroborated this finding in both nations. In contrast, internal political efficacy was positively related to interpersonal trust in the UK but not in Germany.

A high sense of political efficacy should be associated with high subjective well-being (e.g., Whiteley et al., 2010). In line with this assumption, we found that internal and external political efficacy were positively associated with life satisfaction (e.g., Whiteley et al., 2010).

Consistent with the findings of Beierlein, Kemper, et al. (2014b), internal and external political efficacy were independent of left–right self-placement (except for a positive correlation with external political efficacy in the UK).

In agreement with previous studies (e.g., Lane, 1955), internal political efficacy was uncorrelated with authoritarianism in both nations. Although external political efficacy was positively related to authoritarianism in the UK, we found no associations between the two variables in Germany (except for a positive association with the submissiveness subscale).

The correlational pattern of political efficacy with the four justice sensitivity subscales differed considerably across subscales and nations. In the UK, internal political efficacy correlated positively with observer, offender, and beneficiary sensitivity. External political efficacy correlated positively with offender and beneficiary sensitivity. We found no association between political efficacy and justice sensitivity in Germany.

Generally, people tend to bias their answers to self-efficacy questions to appear more favorable to others. In line with this observation, a high sense of (internal and external) political efficacy was related to the exaggeration of positive qualities but also to sincerity when answering items on negative qualities in the UK. In Germany, positive qualities were exaggerated when responding to the internal political efficacy items.

We also calculated correlations between political efficacy and relevant sociodemographic variables—namely employment, income, educational attainment, age, and gender. It has been consistently found that people with a high socioeconomic status have high levels of internal political efficacy (e.g., Beierlein, Kemper, et al., 2014b; Craig & Maggiotto, 1982; Morrell, 2003; Niemi et al., 1991). Our data support this finding: High internal political efficacy was associated with high income, high educational attainment, and male gender in both the UK and Germany, and with high age in Germany only. We further found barely any significant associations between external political efficacy and sociodemographic variables (except for a positive correlation with income in the UK and a negative correlation with age in Germany).

In sum, the pattern of correlations with relevant variables (i.e., the nomological network, see Table 5) confirms the validity of PESS/PEKS in both nations.

Descriptive statistics

Table 6 indicates the reference ranges in terms of means, standard deviations, skewness, and kurtosis of the two subscales of the PESS/PEKS for the total population, as well as separately for gender and age groups in both nations. Standard values are not available.

Table 6

Reference Ranges of the PESS/PEKS Scale Scores for the Total Population and Separately for Gender and Age Groups

−0.75

−0.52

1.17

0.08

−0.48

Note. UK = United Kingdom (N = 468); DE = Germany (N = 474).

Further quality criteria

Due to its short completion time (< 30 s), the PESS/PEKS can be seen as economic.

To investigate to what degree the PESS/PEKS subscales were measurement invariant across the UK and Germany, we estimated multigroup confirmatory factor analysis models (Vandenberg & Lance, 2000; Widaman & Reise, 1997). We tested four successive levels of measurement invariance: configural invariance (same measurement model), metric invariance (same loadings), scalar invariance (same intercepts), and uniqueness invariance (same residual variances). To decide on the achieved level of measurement invariance, we applied Chen’s (2007) cutoff criteria for changes in fit indices. When examining global fit indices, we again applied Hu and Bentler’s (1999) cutoff criteria. We tested measurement invariance based on the essentially tau-equivalent model. For identification purposes, we fixed the first loading and intercept of each factor to one. Due to the identification strategy and our basic model, the configural model equals the metric model. We used MLR estimation to fit the model.

According to these heuristics, the metric model fit quite well (χ²(6) = 30.73, p < .001, CFI = .978, RMSEA = .094, SRMR = .024, BIC = 9729), although the RMSEA was clearly too large. Because the other fit indices pointed to a good model fit, and further modification were not theoretically justified, we accepted the metric model. Thus, latent variances and covariances can be compared across nations (e.g., Bluemke et al., 2016). The scalar model also fit well (χ²(8) = 30.89, p < .001, CFI = .980, RMSEA = .078, SRMR = .024, BIC = 9716). When comparing the scalar with the metric model (Δχ²(2) = 0.44, p = .804, ΔCFI = .002, ΔRMSEA = − .016, ΔSRMR = .000, ΔBIC = − 13), we could accept it according to Chen’s (2007) cutoff values. We even found a lower RMSEA for the scalar compared to the metric model. Scalar invariance implies that latent means additional to latent variance and covariances can be compared across the UK and Germany (e.g., Bluemke et al., 2016). We also accepted the uniqueness model because the global model fit quite well (χ²(12) = 32.28, p < .001, CFI = .978, RMSEA = .067, SRMR = .032, BIC = 9702)[7] and the misspecification induced when restricting the residual variances was negligible (Δχ²(4) = 8.59, p = .072, ΔCFI = − .002, ΔRMSEA = − .011, ΔSRMR = .008, ΔBIC = − 14). Uniqueness invariance implies that researchers working with PESS/PEKS can compare manifest scale scores (means and variances) and correlations across language groups without systematic bias (e.g., Bluemke et al., 2016).

Further literature

The scale was first published in the journal Measurement Instruments for the Social Sciences.

Acknowledgement

We would like to thank Melanie Partsch from GESIS – Leibniz Institute for the Social Sciences for preparing the data. The study was funded by an internal grant provided by GESIS – Leibniz Institute for the Social Sciences, Mannheim, Germany.

Katharina Groskurth, GESIS – Leibniz Institute for the Social Sciences, P.O. Box 12 21 55, 68072 Mannheim, Germany; E-Mail: katharina.groskurth@gesis.org

The dataset supporting the conclusions of this article available in the GESIS SowiDataNet | datorium repository (https://doi.org/10.7802/2081).