Abstract
Most scholars consider the “calling” construct to be multidimensional, yet very little research has examined the dimensions. Of the proposed dimensions, the most unique—and controversial—is a “transcendent summons” toward a particular career. In two studies, we investigated if a transcendent summons uniquely predicts individuals’ endorsement of having a calling, as well as their career-related and general well-being, beyond calling’s other dimensions. Participants were undergraduate students in the U.S. (n = 492) and working adults drawn from a nationally representative, stratified U.S. panel study (n = 767). Results suggested transcendent summons accounted for robust portions of unique variance in perceptions of calling for undergraduates and working adults. Results were mixed for other criterion variables, as a transcendent summons explained variance beyond calling’s other dimensions for three of the five career-related and general well-being variables for undergraduates, and two of five for working adults. Research and practice implications are discussed.
Keywords
Over the last decade-and-a-half, work as a calling has rapidly grown as an area of inquiry in the social sciences. Although the construct may have a “dark side” (Duffy et al., 2016), research has generally demonstrated positive features associated with having a calling (also referred to as “perceiving a calling” or “presence of a calling”) through the construct’s links with numerous well-being criterion variables in work and broader life, including academic satisfaction (e.g., Duffy et al., 2011), job satisfaction (e.g., Xie et al., 2017), meaning in life (e.g., Steger et al., 2010), and many others (Dobrow Riza et al., 2019). Some scholars view calling as unidimensional (see Dobrow & Tosti-Kharas, 2011), but most offer multidimensional conceptualizations (Dik et al., 2012; Hagmaier & Abele, 2012; Praskova et al., 2015; Thompson & Bunderson, 2019; Vianello et al., 2018; Zhang, Herrmann, et al., 2015), with proposed dimensions that vary from one definition to the next. Nearly all definitions include aspects of purposefulness, but “neoclassical” definitions often prioritize aspects of calling’s historical backdrop, including prosocial motivation and a transcendent summons (i.e., a perceived guiding force drawing one toward a particular career path), whereas “modern” definitions emphasize passion or active engagement (Bunderson & Thompson, 2009). Scholars have noted that continued disagreement over calling’s conceptualization may hinder progress in this research area (e.g., Dik & Shimizu, 2019; Thompson & Bunderson, 2019), which makes new research exploring the role and function of calling’s proposed dimensions extremely important.
The transcendent summons dimension can be described as the extent to which a calling is “experienced as originating beyond the self” (Dik & Duffy, 2009, p. 427), and is perhaps the most controversial of its proposed dimensions, given its prominent role among many neoclassical definitions and absence among many modern definitions. Indeed, extant definitions of calling are similar in several respects, but the transcendent summons component falls outside the area of overlap; some oft-cited definitions prioritize it (Dik & Duffy, 2009; Hagmaier & Abele, 2012), while others do not include it (Dobrow & Tosti-Kharas, 2011). Several proposed dimensions of calling (e.g., purposeful work, prosocial orientation, active engagement) evoke other, well-established constructs within vocational psychology and organizational behavior, leading some to question if calling adds anything to what is already known (e.g., Thompson & Bunderson, 2019). Yet the transcendent summons component is unique to calling. In fact, as Brown and Lent (2016) noted, “the transcendent summons element is, arguably, the core and most distinctive aspect of calling” (p. 556). This assertion was made on conceptual grounds, but empirical research also offers a way to evaluate the extent to which a transcendent summons is both unique and a necessary component of calling. One way to examine the dimension’s uniqueness is by investigating its contribution (beyond calling’s other dimensions) in explaining the extent to which people endorse a unidimensional measure of calling. Also, examining the transcendent summons dimension’s role beyond calling’s other dimensions in explaining calling’s well-established relationships with key criterion variables can help further elucidate the importance of the transcendent summons component. The present research sought to explore these research questions using hierarchical linear regression models across two samples.
The Role of a Transcendent Summons
Dik and Duffy (2009) advanced a tripartite conceptualization of calling as a (1) transcendent summons toward (2) purposeful work (i.e., consonance between one’s work and an overarching life purpose) with a (3) prosocial orientation (i.e., a drive toward benefitting others or advancing the common good through one’s work). They argued that a transcendent summons was the defining feature of calling, claiming that those who have discerned purposeful work and a prosocial orientation could be considered as perceiving a “vocation” but not necessarily a calling, which required presence of all three dimensions. Though this distinction between vocation and calling has not been widely adopted in the research literature, it positions the transcendent summons dimension as the factor differentiating calling from other constructs.
Qualitative research has established the transcendent summons dimension’s relevance toward how individuals understand calling across diverse cultural contexts with several studies reporting that this is the most frequently identified aspect of calling. Bunderson and Thompson’s (2009) seminal research on calling among zookeepers from the U.S. and Canada found that many participants attributed the discovery of their calling to a force greater than themselves, such as destiny or the needs of the planet, which are consonant with the “guiding force” notion evoked by a transcendent summons. Likewise, Hunter et al. (2010) found “guiding force” to be the most common theme of calling’s conceptualization among U.S. college students. Furthermore, Hagmaier and Abele (2012) conducted qualitative research on a sample of employed adults in Germany, finding that Transcendent Guiding Force emerged as the most frequently identified theme, accounting for 30.7% of responses. These results led to the creation of a Transcendent Guiding Force dimension on their Multidimensional Calling Measure. A similar process of scale development informed by qualitative research was conducted by Zhang and colleagues (Zhang, Dik, et al., 2015; Zhang, Herrmann, et al., 2015) with 92.4% of their Chinese college student sample’s responses categorized under a “Guiding Force” theme, leading to the creation of their Chinese Calling Scale, which includes a Guiding Force dimension. Notably, this guiding force component has appeared in qualitative findings across samples that represent numerous other cultural and/or occupational contexts, such as Lebanese women (Afiouni & Karam, 2019), ultra-orthodox Israeli women (Goldfarb, 2018), U.S. counseling psychologists (Duffy, Foley, et al., 2012), and undergraduate women at a Canadian Christian university (French & Domene, 2010).
Despite qualitative research indicating that the transcendent summons dimension is crucial to how many people experience a calling, quantitative research on this dimension’s unique role beyond calling’s other dimensions in accounting for calling’s relationship to criterion variables is scarce. In part, this is because most research measuring calling has used total scores, derived either by administering unidimensional scales (e.g., Duffy, Bott, et al., 2012) or by summing the subscale scores of multidimensional instruments (e.g., Lajom et al., 2018). Results from studies that use these research strategies add to our understanding of calling’s overall relationship with criterion variables but do not contribute to a more nuanced, multidimensional understanding of the construct in general nor the transcendent summons dimension in particular.
The lack of research investigating the unique role of a transcendent summons is surprising, given recent calls for such research. In an annual review of vocational psychology scholarship, Brown and Lent (2016) suggested the transcendent summons dimension may be what accounts for the unique predictive value of calling. Similar conclusions were drawn from Thompson and Bunderson’s (2019) effort to synthesize the research on calling within organizational behavior research, which covered 84 articles. Thompson and Bunderson identified both inner requiredness (i.e., fit with personal interests, desires, and sense of meaning) and outer requiredness (i.e., fulfillment of external demand in the world, such as societal needs) as necessary components of what they referred to as a “transcendent calling,” noting that the sense of destiny arrived at when these two coincide is what differentiates calling from a bevy of related constructs. Our review of the literature also suggests the transcendent summons dimension is highly relevant, both within ongoing discourse regarding calling’s definition and within the lives of many individuals across diverse cultural contexts. Quantitative strategies can be used to examine the relevance of this dimension as a unique predictor of key criterion variables, yet no published research (of which we are aware) has investigated this question. For this reason, we sought to investigate the extent to which transcendent summons predicts unidimensional ratings of calling, as well as a series of career-related and general well-being variables. To pursue this goal, we investigated two samples, drawn from populations representing diverse stages of career development among students and workers in the United States (U.S.). In Study 1, we examined these relationships in a sample of undergraduate students, and in Study 2, we conducted analogous analyses in a stratified, nationally representative sample of working adults.
The Present Studies
Because there is little research to draw from regarding transcendent summons’ role in predicting criterion variables, the present study is exploratory in nature. Through this exploratory investigation we sought to examine, in samples of undergraduates and working adults, (1) the extent to which the transcendent summons dimension accounts for unique variance in unidimensional ratings of calling, beyond calling’s other dimensions, and (2) the extent to which the transcendent summons dimension predicts career-related and general well-being variables. Our hope is that results assist in clarifying the importance of the transcendent summons dimension when conceptualizing the construct, and in informing dialogue regarding calling’s operationalization, measurement, and application. Toward this end, we conducted a series of hierarchical linear regression analyses predicting criterion variables in two steps. In Step 1 of each analysis, we predicted criterion variables with only purposeful work and prosocial orientation (i.e., two other frequently-proposed dimensions of calling), and in Step 2, we added the transcendent summons dimension to the model to determine the extent to which it explains additional variance beyond the other dimensions of calling for each criterion variable.
In choosing criterion variables, we began with unidimensional ratings of calling. Including this variable permits an evaluation of the extent to which each specific dimension of calling is necessary in accounting for participants’ endorsement of having a calling. If the transcendent summons dimension does not account for unique variance in unidimensional calling ratings, this would suggest that the transcendent summons may not be a necessary defining characteristic of calling. Next, we focused on several criterion variables previously shown to be associated with overall perceiving a calling, and that were available in the Portraits of American Life Study (PALS; Emerson & Sikkink, 2006-2012) archived data set. Perceiving a calling has been consistently linked to living a calling (Duffy, Bott, et al., 2012), faith-work integration (Neubert & Halbesleben, 2015), academic and job satisfaction (Duffy et al., 2011; Xie et al., 2017), hope (Zhang, Herrmann, et al., 2015), and global appraisals of meaning and purpose in life (Steger et al., 2010), and all of these were available in PALS, so we selected them as criterion variables for analyses in this research. To facilitate parallel data analyses in our two samples, we began by collecting ratings on these same variables from an undergraduate student sample in Study 1. We proceeded to conduct hierarchical regression using perceiving a calling’s dimensions to predict scores on these variables (i.e., unidimensional ratings on perceiving a calling, living a calling, faith-work integration, academic satisfaction, hope, and life meaningfulness). In Study 2, we conducted these same analyses using the PALS dataset—a nationally representative working adult sample—but with job satisfaction replacing academic satisfaction and life purpose in lieu of life meaningfulness.
Study 1
Method
Participants and Procedure
In Study 1, an online survey was administered to an undergraduate psychology student research participation pool from a large, Western U.S. university. A total of 576 participants initiated the survey, but 34 did not proceed beyond the initial few items. These participants were dropped, leaving 542 students who reached the end of the survey. Following Parent’s (2013) recommendations, we retained all participants who completed at least 75% of every scale used in the survey, and calculated participant mean imputation for any item-level missing data. These steps resulted in a sample of 492 participants. Among these, 66.2% self-identified as women and the rest identified as men. The mean age was 19.09 (SD = 1.97) years. Participants identified as White or European (69.5%), Hispanic or Central/South American (11.8%), Multiracial (11.1%), Asian or Pacific Islander (3.8%), Black or African American (1.9%), American Indian or Alaskan Native (.6%), or selected “Other” (1.3%). Participants were mostly in their first year at the university (60.8%), while 24.9% were second-years, 10.5% were third-years, 2.9% were fourth-years, and .8% were “Other.” Regarding socioeconomic status, participants reported growing up in households with estimated incomes of $19,999 or less (4.3%), $20,000 – $34,999 (6.4%), $35,000 – $49,999 (10.0%), $50,000–$64,999 (13.4%), $65,000–$79,999 (16.8%), $80,000–$99,999 (12.1%), and $100,000 or above (37.0%).
Instruments
The below scales were administered (higher scores reflect higher levels of the construct):
Unidimensional presence of a calling
We assessed unidimensional presence of a calling using the Brief Calling Scale (BCS; Dik et al., 2012). The BCS is comprised of four total items that are answered on a 5-point response scale from 1 (not at all true of me) to 5 (totally true of me), two of which can be summed to form a subscale measuring presence of a calling (BCS-Presence). In this study, scores on the two items of this subscale were highly correlated (r = .82). The items of this scale seek to assess calling using face-valid statements (e.g., “I have a calling to a particular kind of work”). Duffy et al. (2015) demonstrated that the BCS-Presence subscale was the strongest predictor among prominent calling measures of participants’ answers to a “yes or no” question asking whether or not they have a calling, suggesting its appropriateness for measuring calling in a way relevant to participants’ personal definitions of calling. BCS-Presence scores have shown strong evidence of construct validity through associations in expected directions with other variables, including life meaningfulness, work hope, and other calling measures (Dik et al., 2012). Thompson and Bunderson‘s (2019) review suggested the BCS is the most widely used measure of unidimensional presence of a calling.
Dimensions of calling
We measured the dimensions of calling using the Calling and Vocation Questionnaire (CVQ; Dik et al., 2012). The CVQ has two broad subscales, CVQ-Presence and CVQ-Search, which measure presence of a calling and search for calling, respectively. These subscales can be further divided into six subscales representing presence and search of Dik and Duffy’s (2009) three dimensions of calling. In this study, we used the 12-item CVQ-Presence subscale to measure the extent to which participants perceive (1) transcendent summons, (2) purposeful work, and (3) prosocial orientation. Participants responded to items using a 4-point scale from 1 (not at all true of me) to 4 (absolutely true of me). Three sets of four items represented transcendent summons (e.g., “I was drawn by something beyond myself to pursue my current line of work”), purposeful work (e.g., “I see my career as a path to purpose in life”), and prosocial orientation (e.g., “Making a difference for others is the primary motivation in my career”). In Dik et al.’s examination of the CVQ’s factor structure, the model acknowledging all dimensions demonstrated acceptable fit, while the model only accounting for unidimensional calling scores showed poor fit. Scores on all three subscales have shown strong internal consistency reliability in student (α = .85–.88; Dik et al., 2012) and working adult (α = .78–.92; Horvath, 2015) U.S. samples. Scores on CVQ subscales have shown construct validity via correlations in expected directions with religious importance, job involvement, and career adaptability, among other variables across diverse cultural contexts (Horvath, 2015; Ponton et al., 2014; Xie et al., 2016). In Study 1, internal consistency reliability was acceptable for purposeful work (α = .79) and prosocial orientation (α = .84) subscale scores. The coefficient for the transcendent summons subscale was lower (α = .65) due to a reverse-scored item. In some studies, researchers have removed this item due to psychometric concerns (e.g., Shimizu et al., 2019), and doing so in our study resulted in a reliable (α = .75) 3-item scale. We conducted all analyses reported in this manuscript using both four- and three-item versions of the subscale. Results were nearly identical (i.e., the only difference between the results with and without the reverse-scored item was that prosocial orientation no longer significantly predicted living a calling without the item), so we reported the results using the standard four-item subscale.
Living a calling
We measured living a calling using the student version of the Living a Calling Scale (LCS; Duffy, Allan, et al., 2012). The LCS is a unidimensional measure that contains six items (e.g., “I have regular opportunities to live out my calling”) assessing the extent to which participants’ current work contexts align with their callings. Theoretically, one cannot live a calling without perceiving a calling (Duffy et al., 2018), so the LCS provides an option to indicate “not applicable – I don’t have a calling.” In line with standard procedure for this questionnaire (e.g., Duffy, Allan, Autin, et al., 2014), we excluded participants who selected the “not applicable” options, resulting in 411 participants for analyses involving this variable. Participants indicated agreement from 1 (strongly disagree) to 7 (strongly agree) for each item. The LCS has validity evidence suggesting that it is an overlapping but separate construct from presence of a calling in student (r = .32; Duffy, Bott, et al., 2012) and working adult (r = .35; Duffy et al., 2013) populations. Likewise, Duffy and colleagues (2011, 2013) have shown that LCS scores positively relate to academic satisfaction, life meaningfulness, and job satisfaction, in addition to mediating relationships between presence of a calling and life satisfaction. In the present study, internal consistency reliability of LCS scores was strong (α = .91).
Faith–work integration
We measured faith-work integration using the 15-item Faith at Work Scale (FWS; Lynn et al., 2009). FWS scores have shown very high internal consistency reliability (e.g., α = .98; Walker, 2013) and were α = .99 in the present study. The FWS assesses how influential individuals’ religious and spiritual beliefs are in their working lives, using items like “I pursue excellence in my work because of my faith,” rated from 1 (never or infrequently) to 5 (always or frequently). The FWS was designed for Judeo-Christian samples, but FWS items have shown applicability to various religious and nonreligious individuals through relationships with job satisfaction and organizational commitment (Neubert & Halbesleben, 2015).
Academic satisfaction
We measured academic satisfaction using Nauta’s (2007) Academic Major Satisfaction Scale (AMSS). Scores on the AMSS have shown strong internal consistency (α = .90; Nauta, 2007), including in the present study (α = .90). Research revealing the AMSS’ capability to predict changes in college majors over two years displays its construct validity (Nauta, 2007). The scale contains six statements (e.g., “Overall, I am happy with the major I’ve chosen”) and uses a 1 (strongly disagree) to 5 (strongly agree) response format.
Hope
We measured hope using the 12-item Dispositional Hope Scale (DHS; Snyder et al., 1991). Four items are unscored “filler” items, and eight items assess hope (e.g., “My past experiences have prepared me well for the future”). Participants assessed how well statements described them from 1 (definitely false) to 4 (definitely true). Snyder and colleagues originally conceived the scale as multidimensional, but later research revealed that unidimensional scores are more appropriate (Brouwer et al., 2008). Patterns of relations among DHS scores with optimism, self-efficacy, and well-being indicate construct validity (Magaletta & Oliver, 1999). DHS scores demonstrated good internal consistency reliability in our study (α = .86).
Life meaningfulness
We used the Meaning in Life Questionnaire (MLQ; Steger et al., 2006) to assess life meaningfulness. The MLQ uses two subscales to assess presence of and search for meaning in life. This study only used the five-item presence subscale (e.g., “My life has a clear sense of purpose”). Items are rated from 1 (absolutely untrue) to 7 (absolutely true) and are aggregated to create a composite life meaningfulness score. Scores on this subscale have shown strong test-retest reliability over the course of one month (r = .70), as well as high internal consistency reliability (α = .81–.86; Steger et al., 2006). The MLQ’s presence subscale has accumulated substantial validity evidence through relationships in expected directions with life satisfaction, depression, anxiety, calling, and numerous other variables (Steger et al., 2006, 2010). In our study, the MLQ showed high internal consistency reliability (α = .87).
Results
Preliminary Analyses
We examined distributions, descriptive statistics, and correlations prior to our main analyses. Weston and Gore’s (2006) guidelines suggest nonnormality exists when absolute values for skewness exceed two and/or kurtosis exceed three. All variables appeared normally distributed, given skewness and kurtosis between −1 and 1. Table 1 provides means, standard deviations, and correlations for all study variables.
Means, Standard Deviations, and Correlations for Study 1.
Note. M and SD are used to represent mean and standard deviation, respectively. * indicates p < .05. ** indicates p < .01.
Hierarchical Linear Regression Models
To examine the unique role of a transcendent summons in calling’s associations with criterion variables, we conducted hierarchical linear regression consisting of two steps designed to isolate the unique contribution of the transcendent summons dimension, as described earlier. Some models demonstrated nonnormality of residuals or heterogeneity, so we followed Field et al.’s (2012) recommendation to account for these conditions through bootstrapping methods. Consequently, an effect was deemed statistically significant when 95% confidence intervals calculated through 5,000 bootstrapped samples did not include zero. Wood (2005) suggested that best practices involve refraining from reporting p-values if the method of determining statistical significance is bootstrapped confidence intervals, and we followed this guidance.
The first model predicted BCS-Presence unidimensional calling scores as the criterion variable. Step 1 regressed BCS-Presence scores on purposeful work and prosocial orientation. In Step 1, significant positive relationships with BCS-Presence were found for both purposeful work (b = .68, CI = .52, .84; β = .42) and prosocial orientation (b = .27, CI = .10, .46; β = .17). Step 1 showed that purposeful work and prosocial orientation together accounted for 29.4% of the variance in BCS-Presence scores. In Step 2, transcendent summons was added and accounted for an additional 6.3% of the variance in BCS-Presence scores beyond that demonstrated in Step 1 (i.e., in total, Step 2 explained 35.7% of the BCS-Presence variance). In Step 2, BCS-Presence scores were significantly positively predicted by transcendent summons (b = .54, CI = .36, .71; β = .32) and purposeful work (b = .44, CI = .23, .63; β = .27). Prosocial orientation no longer served as a significant predictor of BCS-Presence in Step 2 (b = .16, CI = −.01, .34; β = .10).
The next model tested living a calling as the criterion variable. In Step 1, living a calling was significantly predicted by purposeful work (b = .73, CI = .49, .97; β = .38) and prosocial orientation (b = .33, CI = .08, .56; β = .18). These variables accounted for 26.5% of the variance. When added in Step 2, transcendent summons was a significant positive predictor (b = .36, CI = .12, .58; β = .18), explaining 2.4% additional variance in living a calling. Purposeful work (b = .58, CI = .34, .83; β = .31) and prosocial orientation (b = .25, CI = .01, .51; β = .14) remained significant predictors in Step 2.
For Step 1 of the hierarchical regression predicting faith-work integration, purposeful work was a significant positive predictor (b = .23, CI = .03, .40; β = .14) but prosocial orientation was not (b = .05, CI = −.12, .23; β = .03). These variables accounted for 2.6% of the variance in faith-work integration. In Step 2, adding transcendent summons to the model provided an additional 17.6% of explained variance. In Step 2, transcendent summons was the only significant (positive) predictor of faith-work integration (b = .85, CI = .68, 1.01; β = .51), as purposeful work (b = −.15, CI = −.34, .04; β = −.09) and prosocial orientation (b = −.12, CI = −.29, .03; β = −.08) were nonsignificant predictors.
For the hierarchical regression predicting academic satisfaction, purposeful work did not account for unique variance (b = .02, CI −.13, .17; β = .02), but prosocial orientation significantly positively predicted academic satisfaction (b = .31, CI = .16, .46; β = .25). These variables accounted for 7.0% of the variance in academic satisfaction. When we added transcendent summons in Step 2, it explained .2% additional variance and was not a significant predictor (b = .08, CI = −.07, .23; β = .06). Prosocial orientation still significantly predicted academic satisfaction in Step 2 (b = .30, CI = .14, .45; β = .24), but purposeful work was not a significant predictor (b = −.01, CI = −.18, .15; β = −.01).
The hierarchical regression predicting hope revealed significant positive relationships for purposeful work (b = .20, CI = .11, .28; β = .28) and prosocial orientation (b = .11, CI = .03, .18; β = .15) in Step 1. These variables explained 15.8% of the variance in hope. Including transcendent summons added .01% to the explained variance in hope. In Step 2, transcendent summons was not a significant predictor (b = −.01, CI = −.09, .06; β = −.02), while purposeful work (b = .20, CI = .12, .29; β = .29) and prosocial orientation (b = .11, CI = .03, .19; β =.16) remained significant positive predictors of hope.
Finally, in the hierarchical regression predicting life meaningfulness, Step 1 showed significant relationships for purposeful work (b = .39, CI = .18, .58; β = .24) and prosocial orientation (b = .21, CI = .01, .41; β = .13), which accounted for 11.5% of life meaningfulness’ variance. Including transcendent summons in the model added 3.5% explained variance in life meaningfulness. This model revealed that only transcendent summons had a unique (positive) relationship with life meaningfulness (b = .41, CI = .20, .61; β = .25), as associations with purposeful work (b = .21, CI = −.01, .42; β = .13) and prosocial orientation (b = .12, CI = −.08, .31; β = .08) were rendered nonsignificant in Step 2.
Discussion of Study 1
Study 1 first examined the extent to which the transcendent summons dimension accounted for unique variance beyond the other two dimensions of calling (i.e., purposeful work and prosocial orientation) in ratings on a unidimensional measure of calling, with results revealing that it accounted for an additional 6.3%. Next, Study 1 examined the unique influence of a transcendent summons in accounting for established relationships between calling and key criterion variables. Our results established evidence that the transcendent summons dimension explained additional variance in the models examining living a calling, faith-work integration, and life meaningfulness. Indeed, calling’s relationship with the latter two variables was entirely accounted for by the transcendent summons dimension. Our finding that all perceiving a calling dimensions individually contributed to the relationship with living a calling coincides with theory suggesting one must meet the prerequisites of perceiving a calling to live that calling out (e.g., Work as Calling Theory; Duffy et al., 2018). Our results showing that calling’s relationship with faith-work integration was entirely accounted for by transcendent summons is also sensible given scholars’ speculations that the transcendent summons dimension could be particularly important for religious individuals (e.g., Dik, Duffy, & Tix, 2012; Shimizu et al., 2019). Similarly, transcendent summons’ role as the sole unique predictor in calling’s association with life meaningfulness expresses this dimension’s importance in accounting for calling’s well-established link to meaning in life (e.g., Steger et al., 2010), and points to the possibility of an existential longing that may be shared by both a calling and a sense of meaning in life.
Academic satisfaction and hope were not significantly related to the transcendent summons dimension. Prosocial orientation was the only dimension significantly associated with academic satisfaction, while both purposeful work and prosocial orientation variables predicted scores on hope. Hope’s lack of an independent association with transcendent summons may be attributable to the construct’s emphasis on internal states. Snyder and colleagues (1991) noted that hope “is not…defined according to sources external to the person” (p. 571). Perhaps as some experience a transcendent summons as originating from an external source (though certainly not all, see Duffy, Allan, Bott, et al., 2014), the sense of being guided toward a particular career path by something beyond the self is potentially at odds with the internally driven aspects of hope (as defined by Synder et al.), at least in comparison to the other calling dimensions. Bernardo (2010) extended hope theory by differentiating internal locus-of-hope from external locus-of-hope, noting that external locus-of-hope may be partially instilled by the perception of supernatural/spiritual beings or forces. Given this, external locus-of-hope may relate to transcendent summons in ways our study did not show, as the DHS is based on Snyder et al.’s (1991) internally-driven hope conceptualization. Further research into transcendent summons’ associations with hope may benefit from investigating both locus-of-hope factors.
Broadly, Study 1 suggests that the transcendent summons dimension plays a substantial role in informing undergraduate students’ understanding of a calling, and also in calling’s associations with three key criterion variables, even serving as the only unique predictor among calling’s dimensions for faith-work integration and meaning in life. However, transcendent summons was not a factor in predicting scores for academic satisfaction or hope. It was unclear following Study 1 if these results would replicate in other populations, such as working adults whose different career needs may alter the relative influence of calling’s dimensions. This warranted further investigation of our research questions using working adults in Study 2.
Study 2
Study 2 was designed to conduct the same analyses as those in Study 1 but using a national sample of working adults rather than a local sample of undergraduate students. To achieve this, we used data drawn from the Portraits of American Life Study (PALS; Emerson & Sikkink, 2006-2012). Although previous studies have used PALS data with some of these same variables (e.g., Marsh et al., in press), all analyses in the present study are novel. Again, we used transcendent summons, purposeful work, and prosocial orientation for the dimensions of calling. The criterion variables remained largely similar, but job satisfaction was substituted for academic satisfaction to ensure relevance for working adults. Also, we examined life purpose instead of life meaningfulness, as this was the analogous variable available in the archived dataset. This seemed appropriate, as life purpose is considered a facet of life meaningfulness (Martela & Steger, 2016). Again, we used hierarchical linear regression to examine the additive role of the transcendent summons dimension in calling’s association with criterion variables.
Method
Participants and Procedure
This study examined a working adult sample drawn from Wave 2 of PALS, a nationally representative, stratified panel study of adults in the U.S. In 2006, Wave 1 of PALS implemented a four-stage sampling method to collect a nationally representative sample (see Emerson et al., 2010). PALS conducted interviews in home visits, and participants received $50 compensation. Wave 2 of PALS surveyed about half of the original participants and approximately 100 new respondents, totaling 1,419 participants. Most Wave 2 participants completed online surveys (80%) with others interviewed over telephone or in-person. Participants were compensated $30-$50 dependent on interview method. Perry (2016) examined demographic differences between PALS waves, finding no evidence of Wave 2 resulting in biased sample characteristics. Though the PALS data are longitudinal, it is worth noting that items assessing calling were only administered at Wave 2, precluding longitudinal analysis of this study’s research questions. The breadth of the PALS data collection effort was expansive, assessing more than 600 variables. However, the strategic emphasis on breadth came at the expense of severe constraints on the length of individual measurement instruments, given the desire to minimize participant burden. For this reason, we investigated our research questions using single-item and two-item scales.
As the current study exclusively examined full-time and part-time employees, our sample involved 767 participants. The sample was mostly comprised of women (60.9%), and the mean age was 44.44 (SD = 11.92). Regarding ethnicity, the sample self-identified as White (55.0%), Black/African American (16.4%), Hispanic/Latino (15.6%), Asian/Asian American (8.0%), Multiracial (2.7%), Native American (.2%), Pacific Islander (.5%), or selected “Other” (1.4%). Participants reported household incomes as $19,999 or less (9.8%), $20,000 – $34,999 (16.1%), $35,000 – $59,999 (19.7%), $60,000–$79,999 (16.1%), $80,000–$99,999 (11.6%), $100,000 to $149,999 (16.1%), and $150,000 or above (10.5%). 79% of our sample endorsed being full-time workers with 21% identifying as part-time worker.
Instruments
The following brief scales were used. For interpretative ease, we reverse coded some items so higher scores reflected higher levels of each construct.
Unidimensional presence of a calling
Similar to Study 1, the two-item BCS-Presence subscale was used to measure unidimensional calling. The items were highly correlated (r = .77).
Dimensions of calling
Three items were administered to measure the dimensions of calling. The transcendent summons item was “I was drawn by something beyond myself to pursue my current line of work.” For purposeful work, the item “My work helps me live out my life’s purpose” was used. Prosocial orientation was measured using the item “Making a difference for others is the primary motivation in my career.” The second author selected these items from the CVQ because they were judged to best capture the defining characteristics of the relevant constructs. Using CFA on unpublished data in a separate sample, Marsh and Dik (2020) found all items loaded well on their respective underlying factors on the CVQ (λ = .62–.79). Each item used a 5-point response format (“not at all true of me” to “totally true of me”).
Living a calling
Living a calling was assessed using the item “I am living out my calling right now in my job” (1 = “not at all true of me” to 5 = “totally true of me”) selected from the LCS (Duffy, Allan, et al., 2012). This item has been shown to have a strong relationship with total scores on the scale, loading on the latent factor of the six-item LCS at .73 (Marsh et al., in press). It should be noted that the living a calling item in Study 2 lacked a “not applicable” option, precluding us from using exclusion criteria similar to Study 1. Previous research (e.g., Duffy, Bott, et al., 2012) has also used a minimum score cut-off of greater than two on the BCS-Presence subscale to ensure that analyses involving living a calling only include participants who perceive a calling. We conducted analyses on living a calling both using the BCS-Presence cutpoints and not using the cutpoints. These analyses produced very similar results, so we reported the results derived without using the cutoff criteria.
Other criterion variables
Single-item measures were administered to assess faith-work integration (i.e., “I consistently try to live out my faith in my job”; 1 = “not at all true of me” to 5 = “totally true of me”), job satisfaction (i.e., “In general, how satisfied or dissatisfied are you with your job?”; 1 = “very dissatisfied” to 5 = “very satisfied”), hope (i.e., “I feel good about my future”; 1 = “strongly disagree” to 5 = “strongly agree”), and life purpose (i.e., “I believe there is some real purpose for my life”; 1 = “strongly disagree” to 5 = “strongly agree”). Single-item measures have well-established limitations; however, evidence suggests that they can sufficiently capture complex constructs like job satisfaction (Fisher et al., 2016).
Results
Preliminary Analyses
Distributions, descriptive statistics, and correlations were examined prior to main analyses. No study variables exceeded Weston and Gore’s (2006) thresholds for nonnormality. The dimensions of calling were positively correlated with all criterion variables. Means, standard deviations, and correlations for all study variables are reported in Table 2.
Means, Standard Deviations, and Correlations for Study 2.
Note. M and SD are used to represent mean and standard deviation, respectively.
** indicates p < .01.
Hierarchical Linear Regression Models
As in Study 1, we used hierarchical linear regression in two steps to determine transcendent summons’ role above that of the other dimensions. Some linear models had non-normal residual variance or showed visual evidence of heteroscedasticity, so we followed Field and colleagues’ (2012) suggestion to correct for these elements with bootstrapping. We used 95% confidence intervals calculated from 5,000 bootstrapped samples to evaluate statistical significance.
In the first model, we conducted analyses predicting BCS-Presence scores as the criterion variable. Step 1 predicted BCS-Presence scores with purposeful work and prosocial orientation, revealing significant positive associations with BCS-Presence for purposeful work (b = .72, CI = .58, .86; β = .41) and prosocial orientation (b = .54, CI = .40, .68; β = .30). In Step 1, the two predictors explained 40.3% of the variance in BCS-Presence scores. In Step 2, we added transcendent summons to the model, and found that transcendent summons explained an additional 1.9% of BCS-Presence’s variance. Step 2 showed that BCS-Presence scores were significantly positively predicted by all three dimensions of transcendent summons (b = .30, CI = .18, .42; β = .17), purposeful work (b = .61, CI = .46, .75; β = .34), and prosocial orientation (b = .46, CI = .32, .61; β = .26).
Living a calling was examined as the criterion variable in the next model. Step 1 revealed that purposeful work (b = .51, CI = .44, .58; β = .50) and prosocial orientation (b = .29, CI = .21, .36; β = .28) were significant, positive predictors, together accounting for 49.8% of the variance. Adding transcendent summons in Step 2 resulted in an additional .6% of explained variance, and transcendent summons served as a significant positive predictor in this model (b = .09, CI = .02, .16; β = .09). Purposeful work (b = .48, CI = .39, .55; β = .47) and prosocial orientation (b = .26, CI = .19, .34; β = .26) were both significant positive predictors in Step 2.
Next, faith-work integration was examined as the criterion variable. In Step 1, we found significant positive associations for purposeful work (b = .32, CI = .23, .40; β = .30) and prosocial orientation (b = .36, CI = .27, .44; β = .34), together accounting for 32.2% of the variance. When we added transcendent summons in Step 2, the model explained an additional 2.1% of the variance in faith-work integration, and transcendent summons was a significant positive predictor (b = .18, CI = .10, .26; β = .18). In Step 2, faith-work integration was still significantly, positively predicted by purposeful work (b = .25, CI = .15, .34; β = .24), and prosocial orientation (b = .31, CI = .22, .40; β = .29).
In the model predicting job satisfaction, Step 1 showed purposeful work (b = .20, CI = .14, .26; β = .27) was a significant predictor, while prosocial orientation’s relationship with job satisfaction was not significant (b = .00, CI = −.05, .07; β = .00). These variables accounted for 7.1% of job satisfaction’s variance. When transcendent summons was added in Step 2, .2% of additional variance in job satisfaction was explained and transcendent summons (b = −.04, CI = −.10, .02; β = −.05) was not a significant predictor. Step 2 showed purposeful work maintained its significant positive association (b = .22, CI = .15, .28; β = .28), while prosocial orientation was still not a significant predictor (b = .01, CI = −.05, .08; β = .02).
The hierarchical regression model predicting hope revealed significant associations for purposeful work (b = .15, CI = .09, .20; β = .23) and prosocial orientation (b = .06, CI = .01, .12; β = .09), which together explained 8.3% of the variance in hope. Adding transcendent summons to the model in Step 2 did not explain additional variance (0.0%), as transcendent summons was not a significant predictor of hope (b = −.02, CI = −.07, .03; β = −.03). In Step 2, hope was still significantly predicted by purposeful work (b = .16, CI = .09, .22; β = .24) and prosocial orientation (b = .08, CI = .01, .12; β = .10).
Finally, we conducted a hierarchical regression predicting life purpose. In Step 1, we found life purpose to be significantly positively predicted by purposeful work (b = .10, CI = .05, .14; β = .16) and prosocial orientation (b = .14, CI = .09, .18; β = .22), which explained 11.7% of the variance. Step 2’s inclusion of transcendent summons explained .4% additional variance in life purpose. In Step 2, transcendent summons did not significantly predict life purpose (b = .05, CI = −.00, .09; β = .08); however, significant associations were demonstrated for purposeful work (b = .08, CI = .02, .13; β = .13) and prosocial orientation (b = .12, CI = .07, .17; β = .21).
Discussion of Study 2
Study 2 used a nationally representative sample of working adults to investigate the unique role of transcendent summons among perceiving a calling’s dimensions in predicting unidimensional ratings of calling, as well as career-related and well-being variables. In this study, transcendent summons served as a unique predictor of unidimensional perceptions of calling, suggesting that for working adults, as was the case for undergraduates, the transcendent summons dimension is an important element of the way many participants understand their sense of calling. The transcendent summons dimension also was a unique predictor of living a calling and faith-work integration, beyond the other two calling dimensions. However, it is perhaps noteworthy that transcendent summons was the weakest predictor in all of the models in which it accounted for unique variance, which differs from what was found with undergraduate students.
Job satisfaction, hope, and life purpose were all uniquely predicted by purposeful work and/or prosocial orientation but not transcendent summons. This suggests that at least in terms of these variables, working adults may not typically derive substantial improvements through discernment of a transcendent summons beyond what would be garnered through perceiving purposeful work and prosocial orientation. Analyses were not designed to specifically evaluate the unique role of the purposeful work dimension, but it is worth noting that this appeared to be the most prominent dimension among this sample, positively relating to all criterion variable when controlling for the other dimensions. Similar to Study 1, living a calling’s association with all dimensions of perceiving a calling coincides with Work as Calling Theory’s (WCT’s) assertion that perceiving a calling is a necessary precursor to living a calling (Duffy et al., 2018). Faith-work integration’s link to all dimensions in a diverse sample of working adults also reinforces Ponton et al.’s (2014) findings that each dimension was linked to religious importance among employees at a Catholic university.
General Discussion
Responding to scholars’ requests to examine the importance of a transcendent summons (Brown & Lent, 2016; Dik & Duffy, 2009), we conducted hierarchical regression to show the unique role of transcendent summons beyond the other two dimensions of calling, as conceptualized by Dik and Duffy (2009), measured by the Calling and Vocation Questionnaire (Dik et al., 2012), and presented within WCT (Duffy et al., 2018). Study 1 collected data from a U.S. undergraduate student sample, and Study 2 used a nationally representative sample of U.S. working adults. Our results showed, first of all, that a transcendent summons is a necessary component in how undergraduates and working adults conceptualize calling, given it explained unique variance in unidimensional presence of calling ratings beyond calling’s other dimensions.
Transcendent summons did not significantly predict hope nor academic/job satisfaction, but it did have unique associations with living a calling, life meaningfulness (in our student sample), and faith-work integration. Such findings point toward the possibility that the additive role of a transcendent summons may be most relevant to existential concerns and/or spiritual well-being. While Dik et al. (2012) conceptualized transcendent summons as capable of representing sublunary forces (e.g., the needs of society, a family legacy), Hunter et al. (2010) found that many who describe aspects of this dimension tend to cite existential (e.g., “Unique Purpose”) or spiritual/religious (e.g., “Gods Will/Plan”) forces. Such connotations may suggest that transcendent summons’ influence is more germane to variables like life meaningfulness and faith-work integration than toward more motivational or affective experiences such as hope or domain satisfaction. Our mixed findings between Study 1 in which transcendent summons was found to be uniquely associated with life meaningfulness and Study 2 in which it had no unique relation with life purpose has multiple possible explanations. Perhaps this is because life purpose is a narrower construct as only one of the three dimensions of life meaningfulness, representing meaning’s motivational domain (Martela & Steger, 2016). Alternatively, transcendent summons may have a stronger relationship to meaningfulness/purpose among undergraduates than working adults. Differentiating between these explanations in the present research is not possible but represents an interesting and important question for future research to address.
Our findings may help inform the sometimes factious topic of calling’s definition. As has been explored elsewhere (e.g., Thompson & Bunderson, 2019), some definitions of calling highlight aspects of a guiding force beyond oneself (i.e., “neoclassical” definitions; e.g., Dik & Duffy, 2009), while others disregard these aspects, emphasizing personal passion (i.e., “modern” definitions; e.g., Dobrow & Tosti-Kharas, 2011). Transcendent summons showing the strongest relationship to unidimensional calling in Study 1 indicates that the guiding force of neoclassical conceptualizations may hold particular importance for calling’s benefits among students, many of whom may view a career more as a future potentiality than a present reality. In contrast, transcendent summons showing the weakest (albeit still significant) relationship to unidimensional calling scores in Study 2 suggests modern characterizations arguably include the components most relevant to employed adults. This is sensible in light of Dik and Shimizu’s (2019) speculation that different dimensions of calling may vary in their importance between different populations or subgroups within a population, despite empirical results suggesting that there do not appear to be separate “kinds” of callings (Shimizu et al., 2019).
When comparing the strength of relationships across our two studies, it appeared that transcendent summons was a stronger predictor of several criterion variables for students than for working adults. It may be that a transcendent summons is more influential among undergraduate students, an intriguing possibility that raises the question of why this might be so. In Life-Span, Life-Space theory, Super (1980) placed traditional-age undergraduate students in the exploration stage of career development, characterized by developmental tasks that primarily focus on career choice concerns. The notion of a guiding force that can provide direction may seem especially relevant to these concerns, but it may not be as germane to managing tasks in later career stages. This explanation is obviously speculative and requires research designed to directly investigate it, since the present studies did not assess participants’ career stage or related developmental needs. Such research, however, seems promising. For example, research could examine the dimensions’ influences on career development outcomes relevant to exploration tasks among undergraduates to see if a transcendent summons emerges as particularly influential.
Limitations and Future Directions
These studies have several limitations that point to new directions for future research. Both studies are cross-sectional, a design that precludes causal inferences. Longitudinal designs are fortunately becoming more frequent within the literature, and such designs are needed to discern the dynamics that govern how calling’s dimensions influence, and/or are influenced by, various criterion variables over time. Furthermore, the strong intercorrelations among calling’s dimensions suggest that they likely do not operate independently. While transcendent summons’ unique relationship with variables can be statistically disentangled, the processes at work are likely more interactive than these analytic approaches imply.
Study 1 was limited by its homogeneous convenience sample of undergraduate students, although it used well-established measurement strategies. The reverse was true in Study 2, which used much shorter, coarser instruments to collect a very wide but shallow swath of data from a nationally representative sample. These studies’ strengths compensate each other’s limitations, but an optimal future study would use a strong sample and rigorous measurement techniques.
Another limitation of the present study is that it only examined the three dimensions of calling proposed by Dik and Duffy (2009) and adopted by WCT (Duffy et al., 2018). While these dimensions have theoretical basis and are assessed by the most frequently-used multidimensional calling measure, other conceptualizations (e.g., Hagmaier & Abele, 2012; Praskova et al., 2015) propose alternative elements that subsequent studies may investigate. We also only looked at perceiving a calling’s dimensions, but other aspects of calling (e.g., seeking) have a dimensional structure as well; these are worth investigation in future studies. Similarly, our findings on the importance of calling’s dimensions raise questions about whether living a calling, a mechanism through which perceiving a calling elicits positive outcomes (Duffy et al., 2013), may be measured multidimensionally. It is important to note that considerations from theoretical (Duffy et al., 2018) and empirical (e.g., Duffy et al., 2013) work point toward the positioning of living a calling as a mediating variable between perceiving a calling and outcomes. Fleshing out the role of living a calling at the level of calling’s dimensions remains a promising area of future research. To facilitate this research, scholars could create a multidimensional living a calling scale to supplement the already established unidimensional LCS (Duffy, Allan, et al., 2012).
Several psychometric limitations are worth addressing. All measures administered were self-report scales, which makes them susceptible to biases associated with this method (e.g., social desirability bias). One of these concerns regards the potential overlap between the transcendent summons subscale with the instructions of the BCS-Presence subscale. Study 1 administered the BCS alongside its recommended directions (Dik et al., 2012), including the statement, “Broadly speaking, a ‘calling’ refers to a person’s belief that she or he is called upon (by the needs of society, by a person’s own inner potential, by God, by a Higher Power, etc.) to do a particular kind of work.” This may prime individuals to consider a transcendent summons component to a heightened degree. Study 2 did not include these directions, so it is not subject to this particular concern; nevertheless, it is possible that transcendent summons’ stronger relationship with BCS-Presence in Study 1 is at least partially attributable to this aspect. A future study could investigate if Study 1’s findings replicate if the BCS is administered without these instructions. Also, in Study 1, the transcendent summons subscale was less reliable than is typically considered acceptable (α = .65) due to a reverse-coded item. Previous research has found this reverse-coded item to create psychometric limitations (e.g., Shimizu et al., 2019), suggesting revision of the CVQ to optimize its psychometric properties may be fruitful.
Practical Implications
As most studies only show calling’s overall relation to criterion variables, our findings may provide practitioners with more granular guidance. As future research seeking to replicate and expand on these studies accumulates, practitioners can seek to support a calling discernment process for clients around the dimension that tends to be the largest factor in calling’s relationship with the types of outcomes clients are seeking. For example, clients who seek to target existential or spiritual concerns (e.g., finding integration between their faith and their work) may be eager to explore the transcendent summons dimension, whereas clients looking to improve their experience of life or work in more concrete ways (e.g., academic/job satisfaction) through calling discernment may choose to obviate focus on this dimension in favor of establishing a sense of purpose and prosocial impact in their career development.
Our findings also suggested that a client’s life and career status may influence how relevant a transcendent summons is for aiding that client. Generally, students may see larger benefits through discernment of a transcendent summons, while working adults may instead benefit more from focus on cultivating purposeful work. Still, we support calls for practitioners to refrain from treating populations as monolithic groups with identical within-group needs, especially since much remains to be examined regarding the relevance of each dimension based on career status; furthermore, individuals may recycle through previous career stages, whatever their age or employment status (Hartung, 2013).
Finally, results from this study appear to support earlier, theory derived intervention recommendations targeting the transcendent summons dimension. These include assessing the dimension’s relevance for a particular client, framing the discernment process as one tied to existential concerns, steering clients toward active rather than passive strategies of discernment (typically using traditional intervention strategies such as self-assessment, gathering world-of-work information, etc.), and evaluating the possible role of religion and spirituality in career decision-making (e.g., Adams, 2012; Dik & Duffy, 2015; Dik et al., 2009). Research is needed that investigates these strategies specifically, but the studies described in this article help build on a promising foundation of evidence that the transcendent summons is not only conceptually unique, but an established incremental predictor of career-related and general well-being, one worth continuing to investigate.
Footnotes
Authors’ Note
The Portraits of American Life Study (PALS), from which the data for Study 2 were drawn, was led by Michael O. Emerson, currently provost at North Park University. The study was enabled through the generous support of Lilly Endowment Inc. and by supportive funding from Rice University’s Kinder Institute for Urban Research and the University of Notre Dame. For more information, visit 
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Declaration of Conflicting Interests
The author(s) declared no potential conflicts of interest with respect to the research, authorship, and/or publication of this article.
Funding
The author(s) disclosed receipt of the following financial support for the research and/or authorship of this article: This drew a sample of participants from the Portraits of American Life Study (PALS), which received funding from Rice University’s Kinder Institute for Urban Research, University of Notre Dame, and Lilly Endowment Inc.