Engaging Leaders in the Eyes of the Beholder

Abstract

This study investigates whether work engagement mediates the relationship between transformational leadership and service climate. We also examine whether self–other agreement on transformational leadership act as a contingency to the expected relationships. Data were collected from two separate samples. The first sample consists of 1,226 employees of a financial services company while the second sample consists of 291 followers and 30 leaders from an audit company. The results support the view that work engagement mediates the relationship between transformational leadership and service climate. Polynomial regression and interaction analyses show that the relationship between transformational leadership and service climate is moderated by self–other agreement.

Keywords

transformational leadership work engagement service climate self–other agreement

Over the past two decades, researchers and practitioners have stressed the importance of visionary, emotional, and symbolic aspects of leadership (Yukl, 2010). In particular, it has been posited that transformational leadership has the potential to positively influence the well-being of others and organizational and individual performance (Arnold, Turner, Barling, Kelloway, & McKee, 2007). Leaders’ abilities to motivate their followers have been shown to be important in many studies. For example, researchers have documented associations between leadership and follower affect and mood (cf., Gooty, Connelly, Griffith, & Gupta, 2010; Johnson, 2009; Sy, Côté, & Saavedra, 2005), employee motivation (cf., Judge & Piccolo, 2004), and psychological empowerment (cf., Pieterse, Van Knippenberg, Schippers, & Stam, 2010). Although there has been a number of studies of transformational leadership and various motivational/affective variables, there is a relative paucity of studies emphasizing the link between transformational leadership, work engagement, and important organizational outcome variables. The literature search only found three studies that have explored the effects of transformational leadership on work engagement (e.g., Babcock-Roberson & Strickland, 2010; Tims, Bakker, & Xanthopoulou, 2011; Zhu, Avolio, & Walumbwa, 2009), and recent calls have highlighted the need for further exploration on this matter (Bakker, Albrecht, & Leiter, 2011). Clearly, work engagement is an important variable in organizational contexts, and has in recent years been linked to many important work outcomes (cf., Harter, Schmidt, & Hayes, 2002). Thus, the link between transformational leadership and work engagement deserves more attention.

Another important issue is to consider further under which contingencies expected associations between such constructs may hold (Avolio, Walumbwa, & Weber, 2009; Bakker, et al., 2011; Bono & Ilies, 2006). In this regard, both employee characteristics (Zhu et al., 2009) and leaders’ self-evaluations (Atwater, Waldman, Ostroff, Chet, & Johnson, 2005; Atwater, Wang, Smither, & Fleenor, 2009; Atwater & Yammarino, 1992; Heidemeier & Moser, 2009) may serve as contingencies and moderate the proposed associations. In this regard, the degree of self–other agreement in leadership ratings has been seen as an important moderator of the relationship between leadership ratings and outcome variables for some time (Bass & Yammarino, 1991). Further research efforts along these lines are indeed important. Consequently, the first study reported on here discusses and empirically demonstrates how transformational leadership can increase an organisation’s service climate through engendering the positive psychological state of work engagement. The second study shows how an important contextual influence, that of self–other agreement in leadership ratings, may moderate such relationships. Before we proceed on these issues, however, we will discuss the focal concept of work engagement.

Work Engagement

Work engagement has attracted considerable interest from both practitioners and researchers in recent years. Although several definitions and operationalizations exist, there seems to be some agreement that work engagement involves high levels of personal investment in work tasks (Christian, Garza, & Slaughter, 2011). Additional commonalities are enthusiasm, involvement, meaningfulness, and energy (Bakker et al., 2011; Kahn, 1990; Schaufeli & Bakker, 2010; Zhu et al., 2009). The majority of studies of work engagement draw on Kahn’s (1990) conceptual foundation (Christian et al., 2011), which itself draws on the ideas of Maslow (1954) and Alderfer (1972) that people need self-expression and self-employment in their work lives. Kahn (1990) argued that engaged employees apply themselves physically, cognitively, and emotionally. In this view, people differ in how fully they are psychologically present during role performance, which in turn will indicate how engaged they are. Disengaged people, by contrast, are people who withdraw and defend themselves physically, cognitively, or emotionally during task or role performance, which of course can be paralleled with the state of burnout (see Maslach, Schaufeli, & Leiter, 2001). Thus, engagement is a motivational concept that involves the active allocation of personal resources to work tasks (Bakker et al., 2011; Christian et al., 2011; Crawford, LePine, & Rich, 2010; Kahn, 1990; Schaufeli & Bakker, 2010; Zhu et al., 2009). Work engagement can be influenced by another’s experience of engagement indicating potential for social contagion (e.g., Bakker & Demerouti, 2009; Bakker, Van Emmerik, & Euwema, 2006). Thus, work engagement can be studied at the workgroup level of analyses (e.g., Salanova, Agut, & Peiró, 2005). Still, as work engagement represents the personal experiences of individual employees (e.g., Bakker & Leiter, 2010), it has also frequently been studied at the individual level of analysis. One of the most frequently used definitions of work engagement in academia is that of Schaufeli, Salanova, González-Romá, and Bakker (2002). These authors draw on the commonalities of the different conceptualizations of engagement and define it as “a positive, fulfilling, work-related state of mind that is characterized by vigor, dedication, and absorption” (p. 74). Being vigorous refers to having high levels of energy and mental resilience in one’s work as well as being willing to invest effort and persistence. Dedication refers to enthusiasm, inspiration, a sense of significance, pride, and challenge. Finally, being absorbed in work activities refers to having full concentration and being engrossed in one’s work (Salanova & Schaufeli, 2008). The Utrecht Work Engagement Scale (UWES; Bakker et al., 2011; Schaufeli et al., 2002) measures engagement according to the definition of Schaufeli et al. (2002) and includes a subscale for the three dimensions of vigor, dedication, and absorption. Although confirmatory analyses most frequently show that a three-factor structure fit is superior to other factor models, a one-factor structure is also frequently supported and discussed (e.g., Nerstad, Richardsen, & Martinussen, 2010; Sonnentag, 2003). The UWES has been validated in Europe, North America, Africa, Asia, and Australia (Bakker et al., 2011), and it consequently provides solid grounds for assessing work engagement.

Work engagement has been linked to several valuable work-related effects, such as personal health, positive job-related attitudes, extra-role behaviors, job satisfaction, organizational commitment, personal initiative, proactive behavior, learning motivation, and performance (Bakker et al., 2011; Schaufeli & Salanova, 2007). For example, Salanova et al. (2005) found that the levels of the work engagement of contract employees from hotels and restaurants had a significant effect on service quality, as perceived by customers. In addition, Harter et al. (2002) found that levels of engagement were positively related to business-unit performance (i.e., customer satisfaction, profitability, productivity, turnover, and safety).

Given the demonstrated important effects of work engagement, it is crucial to know how to engender it. Previous research has demonstrated that both work-related resources (e.g., Salanova et al., 2005; Salanova & Schaufeli, 2008; Schaufeli & Bakker, 2004) and personal resources (e.g., Salanova, Llorens, Cifre, Martínes, & Schaufeli, 2003; Xanthopoulou, Bakker, Demerouti, & Schaufeli, 2007, 2009) facilitate feelings of work engagement. Beyond this, and as noted above, it is important to show theoretically and empirically how, and under which contingencies, leadership affects employee work engagement.

Transformational Leadership and Engagement

Transformational leadership has been argued to motivate followers to perform beyond expectations (Bass & Bass, 2008). Such leadership should stimulate followers’ interests, create consensus about specific goals, and inspire followers to act in ways that best serve the group and look beyond pure economic self-interest. What is more, transformational leaders have been suggested to raise employees’ needs for self-actualization in the “Maslowian” sense of the term (Bass, 1985). Specifically, transformational leadership is associated with behaviors defined as idealized influence (or charisma) and inspirational motivation, which are displayed when a leader envisions a desirable future and articulates how it can be reached. The leader acts as a role model, sets high standards of performance, and shows determination and confidence. Moreover, transformational leaders influence followers through intellectual stimulation, which is enacted when the leader stimulates creativity and innovation. Finally, transformational leadership is described by individualized consideration, meaning that he or she acknowledges individual differences, develops individuals on their own terms, and qualitatively transforms the reasons for interacting with others from pure self-interest to having interest for others (Bass, 1999). These behaviors have been posited to have individual and organizational consequences, and this has received substantial empirical support. Transformational leadership has been associated with performance ratings at group and organizational levels (Judge & Piccolo, 2004). Followers of transformational leaders have also reported themselves to be more satisfied with their leaders, and by extension with their jobs as a whole, and more strongly motivated to exert extra effort (Bass & Bass, 2008).

Beyond such findings, the increasing interest in positive emotional and affective issues has extended to the leadership domain (Bono & Ilies, 2006). For example, mood contagion has been found to be one of the psychological mechanisms by which leaders influence followers. That is, transformational leaders express more positive emotions, which in turn will have a positive influence on followers’ mood states (e.g., Bono, Foldes, Vinson, & Muros, 2007). In turn, positive emotional expressions and follower mood seem to influence ratings of leader effectiveness and attraction to the leader (Bono et al., 2007; Erez, Misangyi, Johnson, LePine, & Halverson, 2008; Sy et al., 2005). Adding to this, research on work engagement and flow has suggested that optimal experiences cross over from partner to spouse and also from one employee to another (e.g., Bakker, 2005; Salanova et al., 2005). Employees feel engaged because they converge emotionally with the engagement of other members of the group. It also seems that engagement is a collective phenomenon, meaning that some teams or parts of organizations are more engaged than other teams or parts (Salanova et al., 2005). Thus, social psychological group processes seem to be involved in the enhancement and maintenance of work engagement. Therefore, depending on leaders’ capacities to manage such processes, they may have positive impacts on the levels of individual and collective engagement (Schaufeli & Salanova, 2007). Such an influence may again support employee initiative, encourage autonomy, and empower employees to take responsibility along with the provision of increased expertise and interest (Avolio & Bass, 1995). In other words, transformational leaders provide employees with high challenges. Recent research has suggested that high challenges might foster work engagement given that they are perceived as positive by the employee (Crawford et al., 2010; Van den Broeck, De Cuyper, De Witte, & Vansteenkiste, 2010). There is some empirical support for the link between transformational leadership and engagement (Babcock-Roberson & Strickland, 2010; Tims et al., 2011; Zhu et al., 2009). However, this link needs further elaboration with regard to different samples, variables, and contingencies that may affect these relationships. Based on this reasoning and theory, we hypothesize the following:

Hypothesis 1: Transformational leadership is positively related to followers’ levels of work engagement.

Perhaps the most frequently studied effects of transformational leadership are its associations with beneficial job behaviors and job performance. Transformational leaders increase follower task performance and encourage so-called “organizational citizenship behavior,” which is the extra-role behaviors that are not directly recognized by organizations’ formal reward systems even though they typically help improve organizational functioning (Piccolo & Colquitt, 2006). Relationships between transformational leadership and effectiveness are summarized in several meta-analytic reviews (Fuller, Patterson, Hester, & Stringer, 1996; Gang, In-Sue, Courtright, & Colbert, 2011; Judge & Piccolo, 2004; Lowe & Galen Kroeck, 1996). Beyond the consistent findings in this respect, Zohar and Tenne-Gazit (2008) posited that leadership is an antecedent for organizational climate variables. That is, transformational leadership should positively affect the strength of the organizational climate. This association should primarily occur because such leadership is characterized by higher quality leader–member relationships (Wang, Law, Hackett, Wang, & Chen, 2005; Zohar & Tenne-Gazit, 2008). Furthermore, transformational leaders are expected to display greater consistency across situations in terms of their leadership practices, which in turn are expected to reduce variation in group members’ perceptions (Zohar & Tenne-Gazit, 2008). Thus, transformational leaders have effects beyond that of individuals and are expected to influence organizational climate variables. Transformational leadership is also expected to influence the organization’s service climate because of its ability to shape employees’ attitudes toward the customers or end receivers of the organization’s services. In their study, Liao and Chuang (2007) found that managers’ transformational leadership was positively related to employee service performance, which in turn had a positive effect on customers’ loyalty levels. Employee service performance has been highlighted for many years as a vital variable for businesses that experience increased competition, slower growth rates, and mature markets (see Kuenzi & Schminke, 2009; Schneider, White, & Paul, 1998). This is still very much the reality for many businesses, especially viewed in light of the world-spanning financial crisis. Thus, employee service performance is still a vital feature for organizations to engage in. Service climate refers to “employee perceptions of the practices, procedures, and behaviors that get rewarded, supported, and expected with regard to customer service and customer service quality” (Schneider et al., 1998, p. 151). In turn, this climate is proposed to affect employee efforts to deliver service quality to the customers or end users of an organization’s services (cf., Kuenzi & Schminke, 2009; Schneider et al., 1998). Service climate has been positively linked to customer loyalty, customer retention, service-oriented behaviors, perceived service quality, customer satisfaction, and higher organizational performance (Borucki & Burke, 1999; de Jong, de Ruyter, & Lemmink, 2004; Dietz, Pugh, & Wiley, 2004; Rust & Zahorik, 1993; Storbacka, Strandvik, & Grönroos, 1994).

For the reasons described above, we also expect transformational leadership to affect an organization’s service climate. However, as found by Salanova et al. (2005), service climate may additionally be affected by employee work engagement. Thus, our model allows for both the direct and the indirect effects of transformational leadership on service climate. Such partial mediational models allow for other plausible mechanisms that may transmit effects (e.g., de Jong et al., 2004). As these mechanisms are not identified in this model, they are allowed for by the direct effects of transformational leadership on service climate. Based on the above discussion, we hypothesize the following:

Hypothesis 2: Transformational leadership has a positive relationship with followers’ perceptions of the organization’s service climate.

Hypothesis 3: Work engagement partially mediates the relationship between transformational leadership and service climate.

In conclusion, in our first study, we expect a positive relationship between transformational leadership and service climate, which is mediated by work engagement.

Study 1

Method

Sample and Procedure

Data were collected from the employees of a Norwegian financial services company. Using a web-based tool (Confirmit), 3,712 employees received an e-mail with a personalized link to a questionnaire. Of the distributed questionnaires, 1,226 with complete data were returned (33%). The average age of employees was 47 years (SD = 10.5), and 51% were women. The average length of education after high school was 3 years (SD = 2.28), and the average length of employment in the company was 9.5 years (SD = 10.8).

Measures

Transformational leadership

The four dimensions of transformational leadership were measured by an employee report version of the Multifactor Leadership Questionnaire (Bass & Avolio, 1990; we used the transformational leadership scale translated by Hetland & Sandal, 2003). Eight items measured idealized influence, and the other three dimensions (intellectual stimulation, inspirational motivation, and individualized consideration) were measured by four items each, α = .78 to .86. Given the high correlation among the factors (.72-.80), the dimensions were collapsed into one overall transformational leadership variable for descriptive analyses. Furthermore, all four transformational leadership dimensions were positively correlated with both work engagement and service climate (.32-.40, p < .01).

Work engagement

Engagement was measured by a short version of the UWES (i.e., UWES-9; Schaufeli, Bakker, & Salanova, 2006; we used the scale translated by Nerstad et al., 2010). Responses were rated on a Likert-type scale ranging from 1 (never) to 7 (everyday), α = .92.

Service climate

We used the scale developed by Schneider et al. (1998) to measure service climate. Originally, this was a seven-item scale, but we used a shorter version developed by Salanova et al. (2005). The questionnaire was translated into Norwegian by two independent researchers with a back-translation test. Responses were rated on a Likert-type scale ranging from 1 (completely disagree) to 7 (completely agree), α = .66.

Analyses

To test the three hypotheses regarding the relationship between transformational leadership, work engagement, and service climate, we performed structural equation modeling (SEM) analyses using AMOS 18 (Arbuckle, 2009). The method of estimation was maximum likelihood. Because the chi-square difference test is sensitive to sample size, we evaluated model fit based on the overall chi-square/df ratio, Tucker–Lewis index (TLI), comparative fit index (CFI), and root mean square error of approximation (RMSEA; Hu & Bentler, 1998).

When a mediational model involves latent variables, Baron and Kenny (1986) and Judd and Kenny (1981) recommend using SEM. For the sake of parsimony (James, Mulaik, & Brett, 2006), we first tested a model where engagement fully mediated the relationship between transformational leadership and service climate. We then performed additional analyses to test a model specifying only a partial mediational role for work engagement between the independent and dependent variables.

Results and Discussion

The means, standard deviations, and bivariate correlations of the variables are presented in Table 1. The SEM analyses testing our first hypothesis indicated a significant effect of transformational leadership on work engagement while the results bordered on an acceptable model fit (β = .47, p < .001, χ²/df = 13.95, TLI =.93, CFI =.95, RMSEA =.10). We further found a similar effect for transformational leadership on service climate (β = .48, p < .001, χ²/df = 11.26, TLI = .94 CFI = .96, RMSEA = .09). Thus, both our two first hypotheses were supported; however, these restricted models did not fit the data very well.

Table 1.

Descriptive Statistics and Correlations, Study 1.

	M	SD	1	2
1 Transformational leadership	5.04	3.62
2. Work engagement	18.56	0.99	.44**
3. Service climate	4.98	0.97	.39**	.43**

Note. N = 1,226.

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

The results of our mediational models are presented in Table 2. The full mediation research model (M1) was compared with a partial mediation model (M2), which included a direct path from transformational leadership to service climate. As illustrated in Table 2, both models provide satisfactory fit to the data, and M2 provides a somewhat better fit to the data than M1. Because the two competing models are nested, we can use the chi-square difference test (Δχ²) to compare the two models statistically. With a Δχ² of 59.3 and Δdf of 1, M2 is statistically superior to M1 at the .05 level (Hair, Black, Babin, Anderson, & Tatham, 2006). Thus, as postulated in Hypothesis 3, we conclude that engagement partially mediates the relationship between transformational leadership and service climate (see Figure 1).

Table 2.

Model Fit for M1 and M2, Study 1.

	χ²	df	p	χ²/df	TLI	CFI	RMSEA
M1	856.64	88	<.001	9.8	.92	.93	.08
M2	797.34	87	<.001	9.2	.93	.94	.08

Note. TLI = Tucker–Lewis index; CFI = comparative fit index; RMSEA = root mean square error of approximation; M1 = full mediation model; M2 = partial mediation model.

Figure 1.

The final model, Study 1 (standardized path coefficients).

Clearly, the results of Study 1 support our hypotheses that work engagement partially mediates the relationship between transformational leadership and service climate.

Study 2

Our findings in Study 1 supported our hypothesis; work engagement shows its central function in organizational settings as it partly mediates the effects of transformational leadership on service climate. Service climate is an important variable in many settings as it influences customer loyalty, customer retention, service-oriented behaviors, perceived service quality, customer satisfaction, and higher organizational performance (Borucki & Burke, 1999; de Jong et al., 2004; Dietz et al., 2004; Rust & Zahorik, 1993; Storbacka et al., 1994).

Our results, however, may suffer from a few shortcomings and can be nuanced by investigating further how other parameters may influence the relationship between transformational leadership, work engagement, and service climate. For example, the data obtained in Study 1 did not allow us to test the possible moderator effect of self–other agreement or any group effects on the results. This is especially important because both variables have been found to reflect group-level variables in previous research (see Salanova et al., 2005). To investigate this, it is necessary to collect data to calculate the degree of group dependency among scores on the dependent variables (intraclass correration coefficient [ICC] values; see Hox, 2002). In this case, the effects of leadership and work engagement on service climate should be investigated at the aggregate level.

As emphasized in the introductory part of the article, there have been several calls for a more in-depth discussion of how contextual characteristics may influence the effects of leadership. One of the more important of these has been the degree and type of self–other agreement in leadership ratings. The idea is that both self and others’ ratings of leader behaviors may be biased in different ways and that we obtain a more accurate picture of the relationship between leadership and outcome variables when including both types of ratings (Fleenor, Smither, Atwater, Braddy, & Sturm, 2010). Beyond our three hypotheses, we thus explore further if leaders’ self-ratings moderate our expected findings. Despite the consistent promise of transformational leadership affecting both individual and organizational parameters, an increasing number of studies have emphasized the importance of agreement between follower and leader ratings for such predictions. In this regard, Atwater and Yammarino (1997) distinguished between the cases of overrating, underrating, and agreement between self and others’ ratings. Overrating indicates that a person tends to rate his or her performance higher compared with observer ratings. Underrating indicates the opposite. Agreement exists when there is no significant discrepancy between self and observers, beyond the general level of the ratings. Agreement between self and others’ ratings have been described as self-awareness. Fleenor et al. (2010) provided a thorough review of studies addressing those characteristics that affect the degree of congruence between self and others’ ratings. A number of implications of the degree of self-awareness have been identified for both leadership effectiveness and leadership development. Whereas agreement in leader and follower ratings of leadership is related to individual performance and promotability (Atwater et al., 2005), discrepancies are argued to have a severe negative impact (Atwater et al., 2009). Levels of assertiveness and power distance may also moderate self–other agreement (Atwater et al., 2009). Beyond this, the performance implications of over- and underrating seem less pronounced in European settings compared with U.S. settings (Atwater et al., 2005). Thus, since the Norwegian levels of cultural assertiveness and power distance compared with other cultures (see Atwater et al., 2009) are not known, our analyses of self–other agreement on performance have a more exploratory character. That being said, our general expectations are in-line with previous research indicating that overrating is associated with lower levels of outcome variables whereas underrating is associated with elevated levels of outcome variables (see Atwater et al., 2005; Atwater et al., 2009; Fleenor et al., 2010).

A final limitation to our first study is that our data were based on self-reports and may thus suffer from common method bias (Spector, 2006). Unfortunately, none of the post hoc statistical techniques that have been suggested for correcting for such bias has proven to be reliable (Richardson, Simmering, & Sturman, 2009). Thus, the handling of common method bias should rely on other evaluations. As recommended by Conway and Lance (2010), we have made use of validated measures that indicate evidence of appropriate reliability, have a factor structure, and establish relationships with the relevant variables. Another approach to cope with common method variance, and thus to increase the validity of the findings, is to replicate the findings in a different sample (Kline, 2004; Spector, 2006). Replication is additionally important to overcome any problems associated with null hypothesis testing (Kline, 2004).

Taken together, in our second study of the present research we seek to (a) check for statistical dependency between scores (ICC values), (b) replicate the findings obtained in Study 1, and (c) investigate if there is a contingency effect in the degree of self–other agreement on leadership ratings.

Method

Sample and Procedure

Data were collected from the employees of a Norwegian audit company. Responding to recent calls for multiple sources in leadership research (Avolio et al., 2009), questionnaires were distributed to both leaders and employees. Using a web-based tool (Confirmit), all employees (466) and leaders (36) received an e-mail with a personalized link to a questionnaire. Of the employee questionnaires, 291 were completed (62.5%), and of the leader questionnaires, 30 were completed (83.3%). The questionnaires also asked employees to identify the department in which they were employed, which enabled the matching of employee answers to their respective leader’s answers. The minimum number of employees answering by leader was 4 and the maximum was 16. The average age of employees was 42 years (SD = 10.95), and 167 were women. The average number of hours employees spent together with their leaders was 2.7 hours a week (SD = 4.97). The average age of leaders was 49 years (SD = 6.44), and 12 were women.

Measures

The measures for transformational leadership, work engagement, and service climate were the same as the ones used in Study 1, except that transformational leadership was in addition to employee reporting measured by leader self-reporting. Furthermore, employees were asked to report on control variables, such as gender, age, and the number of hours that they typically spent during a week with the leader they reported on. The latter was included because the amount of time spent together may facilitate employees’ leader evaluations. In this second study, the four transformational leadership dimensions were also collapsed into one overall transformational leadership scale because of high correlations among the dimensions (.61-.83 for subordinates) and because they were all positively correlated with both work engagement and service climate (.28-.36, p < .01).

Analyses

Responding to recent calls for multilevel approaches to leadership (Yammarino & Dansereau, 2008), the initial step in our analysis was to check for dependency between scores. This is usually carried out by calculating the ICC values of the dependent variables. However, none of the ICC values of the dependent variables exceeded .04, which has been suggested to be a threshold value for considering scores to be dependent on some source in the data (Heck, Thomas, & Tabata, 2010; Hox, 2002). Thus, in the present context there was too little variation in outcomes between groups to necessitate conducting a multilevel analysis. Instead, we de-aggregated the data so that all analyses could be conducted at the employee level.

To test the three hypotheses we performed the same procedures as in the first study, except for the analyses testing the moderator effects of self–other agreement. For this purpose, we used SPSS 18.0 (IBM SPSS, Chicago, IL) and conducted a series of moderated, hierarchical regression analyses (Cohen & Cohen, 1983). We followed the general advice of Aiken and West (1991) and centered the independent variables before multiplying them with each other. In the moderated regression analysis, the control variables followed by the independent variables were entered and finally the interaction term.

Results

The means, standard deviations, and bivariate correlations of the variables are presented in Table 3. The SEM analyses testing our first hypothesis indicated an acceptable model fit for the relationship between transformational leadership and work engagement (β = .35, p < .001, χ²/df = 1.62, TLI = .94, CFI = .96, RMSEA = .07). We further found a good model fit for our second hypothesis on the relationship between transformational leadership and service climate (β = .44, p < .001, χ²/df = 2.3, TLI = .99, CFI = .99, RMSEA = .02). Thus, both our two first hypotheses were supported.

Table 3.

Descriptive Statistics and Correlations, Study 2.

	M	SD	1	2	3	4	5	6
1. Gender	1.40	0.49
2. Age	41.64	10.95	.064
3. Hours spent with leader	2.70	4.97	−.030	.055
4. Work engagement	4.34	0.91	−.144*	−.108	.083
5. TFL employee report	17.14	3.63	−.088	−.087	.111	.342**
6. TFL self-report	19.20	1.73	.072	−.101	−.055	−.089	.070
7. Service climate	3.54	0.55	.02	−.03	.16*	.27**	.35^**	−.17*

Note. TFL = transformational leadership. N = 291.

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

The results of our mediational models are presented in Table 4. The full mediation research model (M1) was compared with a partial mediation model (M2), which includes a direct path from transformational leadership to service climate. As illustrated in Table 4, M2 provides a somewhat better fit to the data than M1. Because the two competing models are nested, we can use the chi-square difference test (Δχ²) to compare the two models statistically. With a Δχ² of 18.9 and Δdf of 1, M2 is statistically superior to M1 at the .05 level (Hair et al., 2006). Thus, as assumed in Hypothesis 3, we conclude that engagement partially mediates the relationship between transformational leadership and service climate (see Figure 2).

Table 4.

Model Fit for M1 and M2, Study 2.

	χ²	df	p	χ²/df	TLI	CFI	RMSEA
M1	247.24	117	<.001	2.1	.93	.94	.06
M2	228.36	116	<.001	1.5	.94	.95	.06

Note. TLI = Tucker–Lewis index; CFI = comparative fit index; RMSEA = root mean square error of approximation; M1 = full mediation model; M2 = partial mediation model.

Figure 2.

The final model, Study 2 (standardized path coefficients).

To analyze the self–other agreement effects, we conducted a cross-level polynomial regression analysis (Edwards & Parry, 1993; Jansen & Kristof-Brown, 2005). The cross-level polynomial regression analytic technique incorporates polynomial regression within hierarchical linear modeling (Jansen & Kristof-Brown, 2005). Because some of the subordinates shared the same leaders, there is potential noninterdependence (Zhang, Wang, & Shi, 2012). Incorporating with hierarchical linear modeling techniques, this procedure takes into account the potential shared variance to correct for the biases in standard error estimates (Jansen & Kristof-Brown, 2005). Moreover, polynomial regression procedures avoid many shortcomings, such as the reliance on simple statistical techniques apparent in much of the previous difference scores research (e.g., correlation or calculated gap scores; Atwater et al., 2005; Edwards, 1994; Fleenor et al., 2010; Gibson, Cooper, & Conger, 2009). Combined with a response surface methodology, this permits a precise description and evaluation of the component scores (i.e., leader and subordinate ratings) and the relationships studied (Edwards & Parry, 1993). The cross-level polynomial regression equations were as follows:

Level 1 equation:

Z = b_{0} + b_{1} X + b_{2} X^{2} + e

Level 2 equations:

\begin{array}{l} b_{0} = γ_{00} + γ_{01} Y + γ_{02} Y^{2} + μ_{0}, \\ b_{1} = γ_{10} + γ_{11} Y + μ_{1}, \\ b_{2} = γ_{02} + μ_{2} . \end{array}

Z represents the outcome variable of interest (i.e., service climate). X represents subordinate rating on leader transformational behaviors, and Y represents leader self-rating on his or her transformational behaviors.

Response surface analyses are susceptible to outliers (Lambert, Edwards, & Cable, 2003). Therefore, potential outliers were screened using studentized residuals, leverage, and Cook’s D statistics criteria (Fox, 1991; Lambert et al., 2003). Observations that were clearly discrepant from others on the screenplots and exceeded the recommended cutoff values were removed from the analyses. In total, four cases were affected. Table 5 reports the fixed effects estimates of the parameters, including a set of control variables, the two component scores (i.e., subordinate and leader ratings), their squared terms, and their cross products (i.e., interaction terms).

Table 5.

Cross-Level Polynomial Regression Analyses, Study 2.

	Fixed effects coefficients
Variables	Service climate
Control variables:
Subordinate gender	−.17 (.07)**
Subordinate age	−.01 (.00)**
Subordinate education	−.19 (.06)***
Hours spent with leader	.01 (.01)
Leader gender	−.03 (.07)
Leader age	.01 (.01)
Leader education	.16 (.09)*
Component scores:
TFL, employee report (STFL)	.07 (.01)***
TFL, leader self-report (LTFL)	−.07 (.02)***
Squared component scores:
STFL²	.004 (.002)*
STFL × LTFL	−.02 (.01)***
LTFL²	.002 (.01)
Total variance explained	.31***
a₁ (b₁ + b₂)	−.002 (.02)
a₂ (b₃ + b₄ + b₅)	−.009 (.10)
a₃ (b₁ − b₂)	.132 (.02)***
a₄ (b₃ − b₄ + b₅)	.02 (.01)

Note. TFL = transformational Leadership. N(subordinate) = 291, N(leader) = 30. Fixed effects coefficients and their standard errors are shown in each equation. Total variance explained was calculated as 1 – (variance of full model/variance of null model); significance was determined by χ² difference across models. a₁ (b₁ + b₂) and a₂ (b₃ + b₄ + b₅) represent the linear and curvilinear slopes along the congruence line, respectively. a3 (b₁ − b₂) and a4 (b₃ − b₄ + b₅) represent the linear and curvilinear slopes along the incongruence line, respectively.

p < .10. **p < .05. ***p < .01.

The results indicate that service climate was maximized when leaders underrated themselves with a positive coefficient of subordinate rating (b = 0.65, p < .01) and a negative coefficient of leader rating (b = −0.067, p < .01). Moreover, the quadratic term of subordinate rating was positive and marginally significant (b = 0.004, p < .10), indicating a U-shaped curvilinear effect on service climate. In addition, the interaction term of leader–subordinate rating was also significant and negative (b = −0.015, p < .01), implying that leader ratings negatively moderate the positive relationship between subordinate ratings and service climate.

However, to interpret the results, a response surface analysis was conducted (Edwards, 1994) to inspect the linear and curvilinear slopes of both the congruence (X = Y) and the incongruence (X = −Y) axes. The following coefficients were tested and the results are depicted in Table 5:

On the congruence axis, linear slope: a₁ = b₁ + b₂

On the congruence axis, curvilinear slope: a₂ = b₃ + b₄ + b₅

On the incongruence axis, linear slope: a₃ = b₁ – b₂

On the incongruence axis, curvilinear slope: a₄ = b₃ – b₄ + b₅

The results show that both the linear and the curvilinear slopes on the congruence axis (i.e., when a leader and his/her subordinate’s ratings match) were not significantly different from zero (a₁ = −0.002, lower bound = −0.045, and upper bound = 0.041; a₂ = −0.009, lower bound = −0.30, and upper bound = 0.011). This implies that service climate remained at similar levels in this region and that the surface along the congruence line was flat. In other words, regardless of whether leaders and subordinates’ ratings matched at low, medium, or high levels, the degree of service climate was unaffected. By contrast, the linear slope on the incongruence line was significant with a₃ equal to 0.13 and a p value of less than .01. Yet the curvature along the incongruence line was nonsignificant (a₄ = 0.021, lower bound = −0.005, and upper bound = 0.047). Although it was nonsignificant, the slight U-shaped curvature along the incongruence line implied that service climate was less affected when leaders overrated themselves. However, service climate increased when leaders underrated themselves, as shown in Figure 3.

Figure 3.

The response surface analysis, Study 2.

Subsequently, we investigated the anticipated moderation effects. To accomplish this, and since we did not find any acceptable reason to aggregate employees’ leader evaluations, we disaggregated the scores for leaders so that each of their employees’ leader evaluations could be analysed with the leaders’ self-reports using the full sample. In these analyses, we found no significant effects of self-reported transformational leadership on work engagement; nor did we find a significant interaction effect between self-reported transformational leadership and employee-rated transformational leadership on work engagement with or without including the control variables. However, there was a significant linear interaction effect of leader self-report and employee report on service climate (R² change because of the interaction term was .19 (t = 2.12, p = .035), R² for the full model was .22, F(3, 191) =17.61, p < .01; see Table 5.

The plot of this interaction is displayed in Figure 3. Based on a lack of specific hypotheses about the strength of the posited moderated relationship, we plotted our results at several levels of the predictor. These plots showed that leaders who rated themselves lower than their employees (underraters) on transformational leadership had better service climate ratings from their followers. To be certain of our results, we performed a simple effects test as recommended by Aiken and West (1991) to determine whether the slopes were significantly different from zero. These tests showed that the slopes were significantly different from zero at one standard deviation above the mean (simple slope = 0.04, t = 3.08, p = .002) and at one standard deviation below the mean (simple slope = 0.09, t =5.97, p = .000) and lower. The simple slopes were also all positive, indicating that underevaluation was the main cause of the interaction. Taken together, these findings partly support the self–other agreement hypothesis where underraters and overraters are held to perform differently in the leader role (Atwater, Ostroff, Yammarino, & Fleenor, 1998). Finally, we followed a more conservative approach by investigating the interaction, including our three control variables. The results of this analysis indicated that the interaction was also significant when the control variables were included.

Discussion

The two studies reported here respond to recent calls for research on the contingencies under which the links between transformational leadership, work engagement, and performance outcomes occur (Bakker et al., 2011). Our main hypothesis in this respect was supported, as work engagement mediated the relationship between transformational leadership and service climate. Moreover, our expectations regarding self–other agreement were partially fulfilled, and we found that primarily underrating was related to service climate. These results underscore the complexity of leadership research, as the effects of transformational leadership may be contingent on self–other agreement and the degree of employee work engagement. Taken together, the current findings have important implications for both theory and practice, and these are further discussed below.

In both studies, engagement was found to partially mediate the relationship between transformational leadership and service climate. This was expected because of the typically observed climate effects of leadership (e.g., Wang et al., 2005; Zohar & Tenne-Gazit, 2008) and previous findings linking engagement to service climate (Salanova et al., 2005). It thus seems that some of the positive effects of transformational leadership are engendered through the positive state of engagement. This may add to the literature on how transformational, visionary leadership influences employees, and our finding nuances and supports previous ideas that transformational leadership has both affective (Gooty et al., 2010) and motivational (Gooty, Gavin, Johnson, Frazier, & Snow, 2009) implications (see also Ilies, Judge, & Wagner, 2006). Engagement has both affective and motivational components (e.g., Bakker & Leiter, 2010), and it should therefore be expected as a natural consequence of transformational leadership behavior. Specifically, transformational leadership most likely affects employee engagement through an emphasis on vision, intellectual stimulation, and development, and it might thus indirectly influence effort and performance through affect and positive motivational energy. However, transformational leadership also had a weakened but significant independent effect on service climate beyond the effect mediated through engagement.

Interestingly, our findings indicated no significant correlation between leaders’ self-reports and employees’ reports on transformational leadership. Low, or nonexisting, correlations between multiple sources have often been found in previous research and this indicates that individuals will often either overrate or underrate themselves compared with other sources, such as subordinates (e.g., Conway & Huffcutt, 1997). More important though, transformational leadership only seemed to trigger employee work engagement when the leader was perceived as transformational by the employee. This finding was not moderated by leaders’ self-reported transformational leadership and adds to previous research showing similar results, that it is the employees’ understanding of the leader that matters (e.g., Gooty et al., 2009). In conclusion, the perception of engaging leaders is in the eyes of the beholder, and it seems fundamental for employee work engagement that employees themselves perceive their closest leader as being transformational.

When looking into the contingency of self–other agreement, the two methods conducted yielded consistent results. We found that leaders who were modest to low in their self-evaluations received increasing values on service climate with increasing scores on transformational leadership from their employees. Beyond this effect of associated underevaluation, neither did we find any effect of overevaluation nor did we find any effect of the congruence of leader–subordinate ratings in our results. This could imply that leaders that underrate themselves might set more realistic goals for themselves and are more open to feedback than their counterparts, resulting in more positive outcomes (Fleenor et al., 2010). Thus, these results partially correspond to previous research where predictions about leadership outcomes have been found to vary as a function of the level of agreement between leader and subordinate in leadership ratings (Atwater et al., 2005; Atwater & Yammarino, 1992). Our findings in this respect only held for service climate but add to the literature on self–other agreement and cultural influences on the effects of self–other agreement (Atwater et al., 2005; Atwater et al., 2009). In previous studies, the effects of self–other agreement have been found to vary with cultural influences, and such effects have tended to be lower in European than in U.S. settings. Our results showed that the interaction between self and others’ reports was significant in data from our North European culture, but only so that primarily low levels of self-reported transformational leadership, perhaps because of humility on behalf of the leader, produced higher levels of service climate when combined with elevated levels of subordinate-reported transformational leadership. Beyond these findings, the results stress the importance of including multiple sources when evaluating the effects of transformational leadership.

Clearly, these findings have important implications for practice. An increasing number of firms include leadership development programs as part of their personnel strategies. Well as this may be, this study suggests that leaders who think of themselves as transformational do not necessarily have the positive relationships that they anticipate. A clarification of the roles and expectations between leader and follower is potentially an important part of leadership development programs. What is more, several studies suggest that the development of strengths and dispositions (such as increased self-awareness) would positively affect leadership in practice (e.g., Avolio et al., 2009; Avolio & Luthans, 2006; Reichard & Avolio, 2005). Thus, firms would benefit from including both employees and leaders when wanting to enhance leadership skills and leaders’ relationships with their subordinates.

Despite the interesting results of this study, some limitations need to be taken into account. The data collected were cross-sectional, so alternative explanations may exist. It is also possible that there are other causal directions than those indicated in this study. For example, highly engaged employees offering optimal customer service may provide a medium for transformational leadership behaviors. In addition, personality traits could have an important role in the transformational leadership process in such a way that they could make employees more receptive to transformational leadership (Piccolo & Colquitt, 2006).

It should also be taken into account that the data were organized such that each leader response was provided for each leader’s subordinate. Thus, the number of leader responses was increased to make it possible to analyze self–other agreement data. This procedure was carried out because there were no group variables in this study according to our ICC analyses of rater agreement. Thus, future analyses need to test if these results hold when aggregating subordinate responses at the unit level of analysis, which is a more common approach in the transformational leadership literature (e.g., Bass, Avolio, Jung, & Berson, 2003). This nevertheless rests on an assumption of sufficient rater agreement (Heck et al., 2010; Hox, 2002).

This study shows that transformational leadership positively affects service climate through engendering work engagement. Given the interplay between leader reports and employee reports, these findings illustrate the complexity of the leadership phenomenon. Still, this study supports the idea that transformational leadership may contribute to both work engagement and valuable work outcomes. That work engagement also contributes to service climate could indicate that engagement helps create synergy between positive outcomes for employees as well as for organizations. The interplay between these variables indicates that transformational leadership’s relationship with engagement is just as important as its direct effects when creating flourishing organizations.

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 authors received no financial support for the research, authorship, and/or publication of this article.

Author Biographies

Karoline Hofslett Kopperud holds a PhD in organizational behavior and leadership from the Norwegian Business School. Her research interests are related to leadership effects, leader-follower relationality, and positive organizational behavior, especially work-related well-being. She is currently working with leadership consultancy and is involved in research projects and teaching at the Norwegian Business School.

Øyvind L. Martinsen holds a PhD in psychology from the University of Bergen, Norway. He is currently professor in organizational psychology at the Norwegian Business School and adjunct professor at the Norwegian Military Academy. He is member of American Psychological Association and the Association for Psychological Science. His research interests are related to leadership evaluations, personality, creativity, and self leadership.

Sut I. Wong Humborstad is a PhD Candidate in the Department of Leadership and Organizational Behaviour at BI Norwegian Business School. Her research interests include leader-follower relationality, role sending and taking processes, autonomous work practices, and individual proactive work behaviors. Her work has been accepted for publication in journals such as the International Journal of Human Resource Management, Personnel Review, and Chinese Management Studies.

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