validation of a job satisfaction scale in the australian clinical

Validation of aJob SatisfactionScale in the AustralianClinical Medical Workforce

Danny Hills1, Catherine Joyce1 and John Humphreys2

AbstractJob satisfaction has become an increasingly important topic of focus for themedical profession over the last 20 years. This report details the applicationof factor analysis to validate a widely used 10-item job satisfaction scale thathas not previously been validated in a medical practitioner population. Thestudy drew on data from 9,900 participants enrolled in the first wave of a long-itudinal survey of Australian doctors. The instrument was found to possess adominant single factor explaining 75% of the variance and internal reliabilitywas high (r ¼ .86), enabling the determination of a composite job satisfactionscore. Australian doctors experienced high levels of job satisfaction overall,but this varied with doctor subpopulation, age, geographic location, and hoursworked per week. The validation of this brief scale in a large cohort ofAustralian doctors provides opportunities for undertaking further exploratoryand comparative job satisfaction research in medical practitioner populations.

Keywordsjob satisfaction, factor analysis, medical, doctor, work

1 School of Public Health and Preventive Medicine, Monash University2 School of Rural Health, Monash University

Corresponding Author:

Danny Hills, Alfred Hospital Campus, Melbourne 3004, 61399030257

Email: [email protected]

Evaluation & the Health Professions35(1) 47-76

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This report addresses an important concern relating to the psychometric

properties of a job satisfaction scale that has been widely employed in

medical practitioner research over the past 20 years but which has not

been adequately assessed for validity. The primary aim of this study is

to assess the reliability and validity of a 10-item job satisfaction scale, first

adapted for use in a General Practitioner (GP) population (Cooper, Rout,

& Faragher, 1989). The study is undertaken in the context of the first wave

of a large, longitudinal survey of Australian medical practitioners being

conducted annually since 2008. Job satisfaction outcomes from this long-

itudinal survey of Australian doctors are used to demonstrate the applica-

tion of the scale.

Job satisfaction has been an important topic of focus in the organizational,

human resources, and social and behavioral sciences literature. ‘‘Job satisfac-

tion’’ and ‘‘career satisfaction’’ are related constructs, but ‘‘job satisfaction’’

can be differentiated in terms of satisfaction or dissatisfaction experiences

that are specifically job related, as compared to ‘‘career satisfaction,’’ which

is related more to satisfaction or dissatisfaction experiences with an entire

career path (Richardson, Lounsbury, Bhaskar, Gibson, & Drost, 2009). Job

satisfaction may be defined in terms of both pleasure and reward. It comprises

an evaluative judgment of an individual’s job or job situation, reflecting their

responses to the characteristics, challenges, and benefits of the work in which

they are engaged (Weiss, 2002). Thus, job satisfaction can be conceptualized

in terms of ‘‘a related constellation of attitudes about various aspects or facets

of the job’’ (Spector, 1997, p. 2). The key aspects or facets of job satisfaction

have been formally operationalized as ‘‘intrinsic’’ —describing the more

internalized reactions to integral features of the work involved—and ‘‘extrin-

sic’’ —describing features of the job that are more external to the work

involved, such as remuneration, responsibility and autonomy, management

structures, and team relationships (Rose, 2003; Stride, Wall, & Catley,

2007; Warr, Cook, & Wall, 1979; Spector, 2008).

Job satisfaction is important for individuals, organizations, and econo-

mies. It can be considered an important indicator of emotional well-being

or psychological health and how well people are treated at work (Spector,

1997). Job satisfaction has been shown to be strongly related to both mental

and physical health (Faragher, Cass, & Cooper, 2005), at least moderately

related to job performance (Judge, Thoresen, Bono, & Patton, 2001), and

strongly related to organizational commitment (Lok & Crawford, 2004).

Consequently, measuring job satisfaction and its determinants must be a

key consideration for organizations, in terms of understanding workforce

well-being, commitment and productivity, and organizational performance

48 Evaluation & the Health Professions 35(1)



overall. In addition, at the economic and policy level, the measurement of

job satisfaction and its determinants can contribute to the prediction of

future labor market behavior, productivity, and wage models (Long, 2005).

Since the turn of the century, a considerable body of research has been

published in the international literature on medical practitioner job satisfac-

tion across many practice types and specialties, much of it undertaken in

the western, developed countries of North America, Western Europe and

the UK, and Australasia. There is also a growing body of literature report-

ing on medical practitioner job satisfaction in Asian, African, Middle

Eastern, and Eastern European countries. This is not surprising, since job

dissatisfaction in medicine can influence the quality of care provided

(Grembowski, Paschane, Diehr, Katon, Martin, & Patrick, 2005; Landon,

Reschovsky, Pham, & Blumenthal, 2006 Mello et al., 2004; Williams,

Manwell, Konrad, & Linzer, 2007), and decisions to reduce working

hours, to leave clinical practice or to quit medicine altogether (Hann,

Reeves, & Sibbald, 2010; Landon et al., 2006; Rittenhouse, Mertz, Keane,

& Grumbach, 2004; Scott, Gravelle, Simoens, Bojke, & Sibbald, 2006;

Simoens, Scott, & Sibbald, 2002; Williams et al., 2001). Consequently,

the measurement of job satisfaction and its determinants is vitally impor-

tant to ensure that an adequate workforce of committed and competent

health professionals is able to meet the care needs of individuals, commu-

nities, and national populations into the future.

A number of key factors influencing job satisfaction in clinical medicine

have also been reported in the research literature. Income has been consis-

tently identified as a predictor of medical practitioner satisfaction (Cydulka

& Korte, 2008; French et al., 2004; Janus et al., 2008; Leigh, Kravitz,

Schembri, Samuels, & Mobley, 2002; Pratt, 2010; Scott et al., 2006).

Numerous personal, work, and patient-related factors have also been iden-

tified as important drivers of job satisfaction in clinical medicine. While

female gender has generally been found to be related to higher levels of job

satisfaction across many occupational groups in Western countries (Bender,

Donohue, & Heywood, 2005; Kaiser, 2005; Kifle & Kler, 2007; Long,

2005; Sloane & Williams, 2000; Sousa-Poza & Sousa-Poza, 2003), there

have been conflicting findings about this relationship in medicine (Bell,

Bringman, Bush, & Phillips, 2006; Davidson, Lambert, Goldacre, Parkhouse,

& MacDonald, 2002; Lindfors et al., 2007; McGlone & Chenoweth, 2001;

McNearney, Hunnicutt, Maganti, & Rice, 2008; Newbury-Birch & Kamali,

2001; Rosta, Nylenna, & Aasland, 2009). Increasing age, however, has

consistently been shown to be associated with higher job satisfaction in

clinical medicine (Bogue, Guarneri, Reed, Bradley, & Hughes, 2006;

Hills et al. 49



Bovier & Perneger, 2003; Davenport, Henderson, Hogan, Mentzer, Jr.,

& Zwischenberger, 2008; French et al., 2004; McNearney et al., 2008).

Not surprisingly, perhaps, mental health problems have been shown to be

associated with lower levels of job satisfaction in medicine (Dowell,

Hamilton, & McLeod, 2000; Lavanchy et al., 2004; O’Sullivan, Keane,

& Murphy, 2005).

Certain work-related factors have been identified as key drivers of job

satisfaction in medicine. Poorer access to adequate resources and perceived

limitations on the capacity to provide high-quality care, including in rela-

tion to inadequate time, excessive workload, and less cooperative working

relationships, have been shown to be fundamental contributors to lower lev-

els of satisfaction with medical work (Bogue et al., 2006; Janus et al., 2008;

Katerndahl, Parchman, & Wood, 2009; Linzer et al., 2009; Pratt, 2010;

Whalley, Bjoke, Gravelle, & Sibbald, 2006). Higher job stress has been

identified as contributing to lower levels of job satisfaction in medicine

(Grant, 2004; Newbury-Birch & Kamali, 2001; O’Sullivan et al., 2005;

Simoens et al., 2002; Williams et al., 2007) as has working longer hours

(French et al., 2004; Simoens et al., 2002; Ulmer & Harris, 2002). Perceived

autonomy or work control, on the other hand, has been found to contribute

to higher levels of satisfaction in medical work (Bell et al., 2006; Cydulka &

Korte, 2008; Katerndahl et al., 2009; Kinzl, Knotzer, Traweger, Lederer,

Heidegger, & Benzer, 2005; McGlone, & Chenoweth, 2001; McNearney

et al., 2008). Patient-related factors have been shown to be related to lower

satisfaction with medical work, including in relation to the complexity of

care needs (Katerndahl et al., 2009), perceived degree of emotional burden

(Garfinkel, Bagby, Schuller, Dickens, & Schulte, 2005), the threat of legal

action for malpractice (Mello et al., 2004), and community underinsurance

(Pagan, Balasubramanian, & Pauly, 2007). Overall, it appears that job satis-

faction reflects a range of personal, work, and patient-related factors that

may ultimately affect the availability of medical services and the quality

and safety of medical care.

Job satisfaction research is typically undertaken using self-report

questionnaires in cross-sectional or longitudinal surveys of populations or

population samples. Numerous job satisfaction instruments, including

single-item and multi-item scales, have been employed over the last three

decades. While the employment of a single-item measure of global job

satisfaction does confer some psychometric and practical advantages, it is

not considered ideal to jettison the main components of a validated,

multi-item, multifacet scale measuring a complex construct such as job

satisfaction (Martinez-Martin, 2010; Nagy, 2002; Wanous, Reichers, &




Hudy, 1997). Single-item measures of global job satisfaction are

fundamentally biased, since they are most closely correlated with satisfac-

tion related to intrinsic job facets (Rose, 2001) and may enable some

respondents to overstate levels of overall satisfaction by avoiding specific

items that would require the choice of lower ratings that reflect undesirable

personal attributes (Rose, 2003).

The 16-item job satisfaction scale devised by Warr et al. (1979) was vali-

dated in a British population sample of male, ‘‘blue-collar’’ workers

employed full-time in the manufacturing industry and comprises items mea-

suring an intrinsic facet (7 items), an extrinsic facet (8 items), and a single

item measuring job satisfaction overall, with a score for each item ranging

from 1 (extremely dissatisfied) to 7 (extremely satisfied) on a Likert-type

scale. As detailed by Stride et al. (2007), extensive internal reliability and

descriptive statistics have been reported in job satisfaction studies across

industries and occupational groups, and the scale has been found to be sensi-

tive to job control and achievement, role ambiguity and conflict, work pacing

and design, and technological differences and uncertainty. A composite job

satisfaction score is obtained by summing the scores of 15 items, excluding

the overall job satisfaction item, with the option of calculating separate scores

for the intrinsic and extrinsic job satisfaction subscales (Stride et al., 2007).

No other validation studies have been reported.

Variants of the Warr, Cook, and Wall job satisfaction scale have

emerged in the medical research literature, primarily to take account of

the differences between the job characteristics of blue-collar workers in the

original study population and those of tertiary-level trained and primarily

self-employed medical practitioners. The variants include an 8-item short

form utilized with GPs in Ireland (O’Sullivan et al., 2005), a 9-item short

form utilized with GPs in New Zealand and the UK (Appleton, House, &

Dowell, 1998; Dowell et al., 2000; Grant, 2004), and a 10-item short form

utilized with Obstetricians and Gynaecologists in the USA (Bettes, Chalas,

Coleman, & Schulkin, 2004). Each of these studies examined associations

between job satisfaction and other variables but none of the short-form job

satisfaction scales was validated in relation to the population studied. Only

O’Sullivan et al. (2005) reported a Cronbach’s alpha reliability estimate

(r ¼ .83) for the short form utilized. The single item measuring overall job

satisfaction has also been utilized as a measure of job satisfaction in a study

of the effects of pay and job satisfaction on the labor supply of hospital

consultant doctors in Scotland (Ikenwilo & Scott, 2007).

The variant of the Warr, Cook, and Wall job satisfaction scale most com-

monly utilized in medical practitioner research is the 10-item short form

Hills et al. 51



devised by Cooper et al. (1989), which was employed in a study of mental

health, job satisfaction, and job stress experienced by GPs in England.

This 10-item job satisfaction scale was adapted specifically for use with the

GP population. Five intrinsic facet items, 4 extrinsic facet items, and the

overall job satisfaction item were retained from the original scale. Four

extrinsic facet items relating to satisfaction with ‘‘your immediate boss,’’

‘‘industrial relations between management and workers in your firm,’’ ‘‘the

way your firm is managed,’’ and ‘‘your job security,’’ as well as 2 intrinsic

facet items relating to satisfaction with ‘‘your chance of promotion’’ and

‘‘the attention paid to suggestions you make’’ were omitted from the instru-

ment as they were deemed not appropriate for GPs. This determination

would have reflected the largely autonomous, self-employed arrangements

under which the majority of GPs have practiced in the UK. Recent data indi-

cate that over 85% of UK GPs contracted by the National Health Service

remain non-salaried and working either as sole providers or in partnerships

(Technical Steering Committee, 2009).

Total job satisfaction for each respondent was calculated by summing

the scores of the 10 items in the revised scale, and the summed scores were

used as the basis for subsequent descriptive and comparative analyses

(Cooper et al., 1989). The authors did not report on the reliability or validity

of the revised scale but simply reported that test–retest reliability and valid-

ity had been established for the original scale (Cooper et al., 1989). The fail-

ure to validate the revised scale is somewhat surprising. Approximately

30% of scale items were removed from the original 16-item scale to con-

struct the revised scale not on the basis of psychometric concerns per se but

because of the inherent differences between the job characteristics of a sam-

ple of mixed gender, tertiary trained, predominantly self-employed GPs and

the original study sample of male, nonprofessional workers employed full-

time in the manufacturing industry.

Despite the lack of certainty regarding its psychometric properties, the

10-item job satisfaction scale has gained broader acceptance in the medical

practitioner research community and has been employed in a range of

medical practitioner job satisfaction studies, internationally. It includes

key facets of job satisfaction relevant to medical practitioners and can be

considered acceptable for use with self-employed or salaried clinicians,

academic medical practitioners, or those engaged in mixed-sector employ-

ment. On this basis, the scale may be considered to possess ‘‘face validity,’’

although this can be legitimately viewed only as a starting point for more

rigorous testing (Downing, 2006; Sartori, 2010). The brevity of the scale

is also likely to make it an attractive option for researchers attempting to




obtain adequate response rates in medical practitioner surveys (VanGeest,

Johnson, & Welch, 2007).

Most applications of the 10-item job satisfaction scale in research on

medical practitioner populations have been conducted in the UK, primarily

in relation to job satisfaction and stress sources or symptoms in GP cohorts

(Rout, 1999; Rout & Rout, 1994; Sibbald, & Bojke, 2003; Sibbald, Enzer,

Cooper, Rout, & Sutherland, 2000; Sutherland & Cooper, 1992, 1993;

Whalley et al., 2006). Other UK studies have used the scale to investigate

job satisfaction and intentions to retire or otherwise leave medical practice

(French et al., 2004; Scott et al., 2006; Sibbald, Bojke, & Gravelle, 2003;

Simoens et al., 2002). The scale has also been used to estimate job satisfac-

tion among doctors in Norway (Aasland, Rosta, & Nylenna, 2010; Nylenna,

Gulbrandsen, Førde, & Aasland, 2005a, 2005b), hospital doctors in Norway

and Germany (Rosta et al., 2009), rural physicians in Canada (Lavanchy

et al., 2004), and GPs in Australia (Harris et al., 2007; Ulmer & Harris,

2002; Walker & Pirotta, 2007).

Importantly, in subsequent studies utilizing the 10-item job satisfaction

scale in medical practitioner populations, no attempts to validate the scale

have been reported. Only Rout (1999) reported an internal reliability esti-

mate (r ¼ .82) for the revised scale. Furthermore, the scoring methods used

across different studies have been inconsistent and have included calculat-

ing a composite score by summing all items or calculating a mean of the

item scores (Harris et al., 2007; Nylenna et al., 2005a, 2005b; Sutherland

& Cooper, 1993; Walker & Pirotta, 2007), utilizing the mean or median for

each item (French et al., 2004; Lavanchy et al., 2004; Rosta et al., 2009;

Rout, 1999; Rout & Rout, 1994; Scott et al., 2006; Sibbald et al., 2000;

Sutherland & Cooper, 1992; Ulmer & Harris, 2002), or reporting the mean

for each item but primarily utilizing the single, overall job satisfaction item

in analyses (Sibbald & Bojke, 2003; Sibbald et al., 2003; Simoens et al.,

2002; Whalley et al., 2006).

Given the lack of clarity about the psychometric properties of the

10-item job satisfaction scale, it is not surprising that a limited number of

studies employing the scale have reported a composite job satisfaction score

as the basis for descriptive and comparative analysis. If the scale was deter-

mined to be a single-factor scale, a composite score of the total job satisfac-

tion construct could be determined for each respondent by summing the

scores of individual items, calculating a mean score for all items or produc-

ing a standardized score (Petit, Lackey, & Sullivan, 2003). The advantage

of obtaining a composite score for a construct is that it can be more easily

interpreted, can be used to assess differences between groups of

Hills et al. 53



respondents, and can be compared across studies (Petit et al., 2003).

Obtaining composite scores for different groups in a population or population

sample can also facilitate the establishment of comparisons or relationships

with other key variables and the development of patterns of characteristics

or performance in the population or population sample (Child, 2006).

Consequently, the primary purpose of the current study was to assess the

reliability and validity of the 10-item job satisfaction scale that has been

employed in at least 20 medical practitioner studies in the UK, Europe,

North America, and Australasia over the last two decades, including a large

longitudinal study of Australian doctors that commenced in 2008. Estab-

lishing the validity of the scale underpins the viability of future applications

of the scale in job satisfaction research in medical practitioner populations,

as well as any consideration of employing the scale more broadly. In addi-

tion, in order to demonstrate the practical application of the scale, a descrip-

tive analysis of job satisfaction outcomes for a large cohort of Australian

doctors was conducted.

Method

Data collected in the first wave of a prospective cohort study of Australian

doctors engaged in clinical medical practice in 2008 were utilized for this

validation study. The Medicine in Australia: Balancing Employment

and Life (MABEL) survey is a national, longitudinal study investigating the

patterns and determinants of clinical medical workforce participation in

Australia and has been designed to obtain empirical evidence to support

policy responses to a range of medical workforce issues. The survey is

being conducted annually until at least 2011. The sampling frame for the

survey is the Australian Medical Publishing Company Medical Directory,

a national database of Australian medical practitioners, which is regularly

updated and used extensively for mailing purposes. Doctors in Australia

work in a variety of clinical settings either in public or private sector prac-

tice or a combination of both.

Four customized questionnaires were developed, specifically tailored for

each of the four subpopulations of GPs and GP registrars, specialists, spe-

cialists in training, and hospital non-specialists. Each questionnaire could

be completed either online or in hard copy. All of the 54,750 Australian

doctors engaged in clinical medical practice in 2008 were invited to partic-

ipate in the MABEL survey. The conduct of the study was approved by the

University of Melbourne Faculty of Economics and Commerce Human

Ethics Advisory Group and the Monash University Standing Committee




on Ethics in Research Involving Humans. A detailed account of the

MABEL study protocol and baseline data has demonstrated that respon-

dents to the baseline survey in 2008 comprised a representative cohort of

Australian doctors (Joyce et al., 2010).

All questionnaires in the first wave of the MABEL survey included the

10-item revised job satisfaction scale devised by Cooper et al. (1989) but

with slightly modified item response options. This scale was selected

because there was some evidence of reliability that could be retested in the

MABEL study population, and it was considered to reflect the key aspects

of job satisfaction relevant to the autonomous, self-employed working

arrangements of the majority of medical practitioners in Australia. Conse-

quently, it was expected that the MABEL study findings would provide new

information about job satisfaction in Australian clinical medical practice

and be more readily compared with the results of studies conducted in sim-

ilar populations. On a practical level, the scale is brief enough to be

included in the annual survey questionnaire with a large range of other vari-

ables that are likely to be associated with workforce dynamics in Australian

clinical medical practice.

Respondents were asked to indicate how satisfied or dissatisfied they

were with each of the various aspects of their work as a doctor represented

by the scale items. Scale items and the indication of intrinsic or extrinsic

facets of work, as originally determined by Warr et al. (1979), are listed

in Table 1. The original 7-point Likert-type item scales were reduced to

5-point item scales, each scored as 0 ¼ Very dissatisfied, 1 ¼ Moderately

dissatisfied, 2 ¼ Not sure, 3 ¼Moderately satisfied, and 4¼ Very satisfied.

Table 1. Job Satisfaction Scale Items

Item Intrinsic Extrinsic

1. Freedom to choose your own method of working �2. Amount of variety in your work �3. Physical working conditions �4. Opportunities to use your abilities �5. Your colleagues and fellow workers �6. Recognition you get for good work �7. Your hours of work �8. Your remuneration �9. Amount of responsibility you are given �10. Taking everything into consideration,

how do you feel about your work?

Hills et al. 55



A Not applicable response option was also provided for each item. The

number of item response options was reduced primarily for reasons of space

and to maintain a degree of uniformity in item response options for the con-

siderable number of item scales in each of the MABEL survey instruments.

Cleaned data were stored in a Stata (StataCorp, 2009a) database and sub-

sequently extracted for analysis. All data analyses were undertaken in Stata,

Release 11 (StataCorp, 2009a), which included the polychoric correlation

add-on (Kolenikov, n.d.). Each of the job satisfaction variables was

reviewed and blank, unreadable, and not applicable responses were all

recoded to missing values in order to facilitate analysis. To detect any

potential selection bias between respondents with missing data in the job

satisfaction variables and respondents without missing data, a comparison

was undertaken using Pearson’s chi-squared test of independence in relation

to doctor subpopulation, age, gender, state of practice, rurality of practice

location, and hours of work per week. Since the item scales provided only

ordinal-level data, polychoric correlation was the most appropriate and

more accurate measurement model to use in factor analysis (Gilley & Uhlig,

1993; Holgado-Tello, Chacon-Moscoso, Barbero-Garcıa, & Vila-Abad,

2010; Pearson & Pearson, 1922). To facilitate comparison of outcomes

using alternative correlational methods, however, factor analytic proce-

dures were replicated with a nonparametric (Spearman’s rho) correlation

matrix, as the distribution of responses for each job satisfaction variable was

determined to be highly skewed.

Inter-item correlational analysis was undertaken listwise to eliminate

respondent sets of data in which there were one or more missing coded val-

ues, which ensured subsequent analyses could be conducted with the same

population proportion data set. The significance of each of the correlations

was reviewed to ensure that there were sufficient significant correlations

to justify undertaking exploratory factor analysis (Petit et al., 2003). In

addition, a visual inspection of the correlation matrices was undertaken to

determine the presence of any highly correlated pairs that may represent

item duplication. The inbuilt ‘‘factortest’’ command in Stata (StataCorp,

2009a) was utilized to compute the determinant of the correlation matrix,

Bartlett’s test of sphericity, and the Keyser-Meyer-Olkin (KMO) measure-

ment of sampling adequacy (Petit et al., 2003).

Exploratory factor analysis of each of the correlation matrices was

undertaken in five primary steps. First, each matrix was subjected to Iter-

ated Principal Factor (IPF) analysis in Stata, which is equivalent to Princi-

pal Axis Factoring analysis. The second step involved the extraction of

nontrivial factors and determining the need for the rotation of factors. In the




current analysis, eigenvalues of the factor solutions and screeplots were

examined for each matrix. The third step comprised assessing the factor

loadings of scale items. Factor loadings less than 0.3 were not retained

since, at this level, less than 10% of the item variance (the square of the

loading) is accounted for by the factor. Reflecting the large sample size and

the relatively small number of scale items, this cutoff point would ensure

that loadings have practical significance (Hair, Black, Babin, & Anderson,

2010). The fourth step involved assessing the internal reliability of the fac-

tored scale, using Cronbach’s coefficient alpha on the unstandardized item

scores. Finally, to test the replicability of the subsequent factor structure

throughout the subpopulations of Australian clinical medical practitioners,

IPF analysis was also undertaken on the data sets for each of the four sub-

populations of doctor.

With the establishment of a factor structure, secondary descriptive anal-

yses were undertaken on job satisfaction outcomes for the population of

Australian doctors in relation to the key profile variables of doctor type

(subpopulation), gender, age, geographic location by state and rurality, and

hours worked per week. ‘‘Rurality’’ was defined in terms of the Australian

Standard Geographical Classification (ASGC) remoteness areas system,

which classifies locations into the comparative categories of ‘‘major city,’’

‘‘inner regional,’’ ‘‘outer regional,’’ ‘‘remote,’’ or ‘‘very remote’’ based on

an index of population size and accessibility (Australian Bureau of Statis-

tics, 2003; Australian Institute of Health and Welfare, 2004).

Results

The first wave of the MABEL survey established a baseline cohort of

10,498 Australian doctors, representing 19.36% of all Australian doctors

engaged in clinical medical practice in 2008 (N ¼ 54,750). Respondents

comprised 3,906 GPs and GP registrars, 4,596 specialists, 924 hospital non-

specialists, and 1,072 specialists in training, which provided a representa-

tive cohort of the Australian clinical medical workforce with only minor

response biases detected in relation to doctor subpopulation, gender, age,

geographic location, and hours worked each week (Joyce et al., 2010).

Response patterns for the job satisfaction scale items in the present study

were negatively skewed, with median scores of 3 in a range from 0 to 4,

which demonstrated that the majority of respondents were satisfied with

each aspect of their work represented by the scale item. The lowest levels

of satisfaction were recorded for Item 6 (’’recognition you get for good

work’’), Item 7 (‘‘your hours of work’’), and Item 8 (‘‘your remuneration’’),

Hills et al. 57



with 64.6–67.4% of respondents satisfied or very satisfied with that aspect

of their work. Over 90% of respondents indicated that they were satisfied or

very satisfied in relation to Item 2 (‘‘amount of variety in your work’’). For

the remaining items, 83.4–87.9% of respondents were satisfied or very sat-

isfied with that aspect of their work.

Respondents delivering a missing value in at least one of the 10 job

satisfaction variables comprised 5.29% (555) of the baseline cohort and

were excluded listwise from factor analysis, with the remaining 94.71%(9,943) included in analyses. Comparisons of included and excluded

respondent profile variables and the significance of differences in the vari-

ables were determined by the Pearson chi-squared test of independence.

Profile variables comprised doctor type, gender, age, geographic location

by state and rurality, and hours worked each week. Overall, there were min-

imal differences between the profile of the 9,943 respondents whose data

were included in factor analysis and those whose data were excluded. There

were no significant differences between the groups in terms of gender and

location of practice by state or rurality; however, the excluded group com-

prised a slightly greater proportion of GPs and a slightly reduced proportion

of specialists in training were more likely to be older and were more likely

to work fewer hours per week.

The polychoric and Spearman’s rho inter-item correlations were deter-

mined for all included respondents. All correlations were found to be signif-

icant (p < .00001). The Spearman’s rho correlations were uniformly lower

in magnitude than the polychoric correlations, but the patterns of associa-

tion were very similar. The lowest levels of association in both matrices

were detected between both Item 7 (‘‘your hours of work’’) and Item 8

(‘‘your remuneration’’) and each of Item 2 (‘‘amount and variety of your

work’’) and Item 5 (‘‘your colleagues and coworkers’’). The strongest asso-

ciation in both matrices was between Item 1 (‘‘freedom to choose your own

method of working’’) and Item 10 (‘‘taking everything into consideration,

how do you feel about your work?’’).

Visual examination of the correlation matrices suggested ‘‘factorabil-

ity,’’ as did other analytic outcomes, including the determinant of the matrix

of correlation (.029), Bartlett’s test of sphericity (X2¼ 35260.5, df¼ 45, p <

.001), and both the pre-estimation and post-estimation determinations of the

KMO measure of sampling adequacy (0.91). While the low determination

value reflects the relatively strong associations (>0.6) between some vari-

ables, the significant outcome for the test of sphericity and the very high

measures of sampling adequacy provides strong support for the ‘‘factorabil-

ity’’ of the matrices (Petit et al., 2003).




IPF analysis of both the polychoric and Spearman’s rho correlational

matrices resulted in initial, nine-factor solutions. A dominant first factor

exhibited a large eigenvalue for the both the polychoric (5.01) and

the Spearman’s rho (4.11) solutions, explaining approximately 75% of the

variance in relation to the total variance. Examination of the screeplots

(Figure 1) demonstrates the primacy of the first factor and the presence

of a relatively trivial second factor just above the ‘‘scree.’’ A single-

factor solution satisfied both the Kaiser criterion for retaining factors with

an eigenvalue greater than 1.0 and the Joliffe criterion for cropping factors

with an eigenvalue less than 0.7 (Lance, Butts, & Michels, 2006).

The factor loadings produced by IPF analyses demonstrated a similar

pattern in both the polychoric and the Spearman’s correlational matrices.

All scale items had high loadings on the first factor, ranging from 0.57 to

0.92 for the polychoric and 0.49 to 0.84 for the Spearman’s rho matrices.

The first factor thus accounted for between 25 and 85% of the variance

in each item. Duplicate loadings greater than 0.3 in the lesser factors were

detected for Items 2, 4, 7, and 8 in Factor 2 and Items 5 and 6 in Factor 3, but

these were generally at least 0.2 in absolute magnitude less than the load-

ings in the first factor. Communality was moderate to high for each scale

item, ranging from 0.51 to 0.88 in the polychoric and 0.42 to 0.74 in the

Spearman’s rho matrices, reflecting the level of shared variance among the

scale items. These outcomes further supported the determination of a

single-factor solution comprising all scale items. Consequently, no attempt

was made to rotate the factor solution.

The final stage of factor analytic procedures comprised the completion

of internal reliability analysis of the scale, determining Cronbach’s alpha

coefficient in relation to the unstandardized item scores (Table 2). The

item–test correlation output shows the pairwise correlation between each

item and the whole scale, whereas the item–rest correlation output shows

the pairwise correlation between each item and the remaining items in the

scale (StataCorp, 2009b). The average inter-item covariances were similar

across all items and the subsequent alpha for the scale was high (.86).

Removing any item would have reduced the overall alpha of the scale, as

indicated in the final column, further strengthening the case for a single-

factor solution for the 10-item job satisfaction scale.

Factor analytic procedures were replicated for each of the four MABEL

survey doctor subpopulations of GPs and GP registrars, specialists, hospital

non-specialists, and specialists in training. Similar outcomes to that for all

doctors were obtained. KMO measures of sampling adequacy remained

high, ranging from 0.89 to 0.92. IPF analyses identified a strong first factor,

Hills et al. 59



Polychoric

0.0

1.0

2.0

3.0

4.0

5.0

6.0

1 2 3 4 5 6 7 8 9

Eigenvalue

Factor

2 3 4 5 6 7 8 9

Factor

Spearman’s Rho

0.0

1.0

2.0

3.0

4.0

5.0

6.0

1

Eigenvalue

Figure 1. Iterated principal factor analysis screeplots.




with eigenvalues ranging from 3.7 to 5.2, representing 70–77% of variance

in relation to the total variance, and a minor second factor with eigenvalues

ranging from 0.60 to 0.87 representing 10–14% of variance in relation to the

total variance. Duplicate loadings in other factors were most commonly

detected for Items 2, 4, 7, and 8, each loading at lower levels than in the first

factor. Internal reliability analyses for each doctor type maintained consis-

tently high Cronbach’s alpha coefficients, ranging from 0.83 to 0.86, with

item removal identified as reducing the overall scale alphas.

Based on the single-factor solution for the 10-item scale, a total job satis-

faction score was determined for each respondent to the MABEL survey

scoring all 10 items in the scale (N¼ 9,943), in the range of 0–40 (Figure 2).

In a significantly left-skewed (�0.96) and kurtotic (4.26) distribution of

total job satisfaction scores, and reflecting individual scale item responses,

Australian doctors were found to experience high levels of job satisfaction,

with 50.7% scoring greater than 30 and 90.1% scoring greater than 20

(M ¼ 29.7, SD ¼ 6.90). Although distinctly non-normal, the distribution

demonstrates a smooth curvilinear shape approaching the mean with an

observable degree of discrimination across the range of scores, including

at the higher levels of job satisfaction. This outcome is consistent with the

levels of job satisfaction that would be expected from a professional popu-

lation that is highly educated, well remunerated, and generally experiences

high levels of professional autonomy and status.

Due to the non-normal distribution of total job satisfaction scores and

violations to the assumption of equal variances for the profile variables,

Table 2. Internal Reliability Analysis (N ¼ 9,943)

ItemItem–test

correlationItem–rest

correlationAverage inter-item

covariance Alpha

1 .7137 .6288 .4009 .84082 .6106 .5245 .4320 .84993 .6302 .5194 .4125 .85034 .6934 .6101 .4095 .84285 .5546 .4566 .4392 .85446 .7124 .6135 .3907 .84207 .6274 .5004 .4057 .85358 .6431 .5152 .3999 .85279 .6862 .6036 .4126 .843510 .8339 .7848 .3866 .8294Test scale .4090 .8593

Hills et al. 61



as determined by significant Bartlett’s tests for equal variances (p < .05),

and the failure of transformations of total job satisfaction scores to approach

normality, the Kruskal-Wallis equality-of-populations rank test (Hamilton,

2009) was employed to test for the equality of median job satisfaction

scores between profile variable categories. As indicated in Table 3, signif-

icant differences in median job satisfaction scores were detected for doctor

type, age group, geographic location by state and rurality, and hours worked

each week but were not detected for gender.

With regard to doctor type, lower median total job satisfaction scores

were detected for hospital non-specialists (27) and specialists in training

(28) with higher median scores detected for GPs (31) and specialists (31).

Doctors aged younger than 35 years had the lowest median total job satis-

faction score (28), whereas doctors aged 65 years and older had the highest

median total job satisfaction score (35). Although the Kruskal-Wallis test

detected significant differences in total job satisfaction in relation to respon-

dent location by state, it is difficult to discern differences by simply review-

ing the mean or median scores. In comparison, median total job satisfaction

for doctors in very remote locations (32.5) was distinctly higher than for

doctors in other geographic locations (31). In an apparent ‘‘dose–response’’

relationship between hours worked and decreasing median total job

0

100

200

300

400

500

600

700

0 5 10 15 20 25 30 35 40

Freq

uenc

y

Total job satisfaction score

Figure 2. Distribution of total job satisfaction scores (N ¼ 9,943)




Tab

le3.

Tota

lJo

bSa

tisf

action

Score

s

Tota

lJo

bSa

tisf

action

Score

sK

rusk

al-W

allis

Tes

tPro

file

Cat

egory

n(%

)R

ange

MSD

Mdn

X2

(Sig

nifi

cance

)

Doct

or

type

(N¼

9,9

43)

GPs

and

GP

regi

stra

rs3,6

72

(36.9

3)

0–40

30.3

56.7

931

406.8

65

(p<

.001)

Spec

ialis

ts4,3

58

(43.8

3)

0–40

30.2

66.8

831

Hosp

ital

nonsp

ecia

lists

883

(8.8

8)

3–40

26.5

46.8

627

Spec

ialis

tsin

trai

nin

g1,0

30

(10.3

6)

2–40

27.8

16.3

228

Gen

der

(N¼

9,9

42)

Mal

e6,0

20

(60.5

5)

0–40

29.6

07.0

531

1.1

13

(p¼

.291)

Fem

ale

3,9

22

(39.4

5)

0–40

29.8

86.6

531

Age

(N¼

9,8

64)

<35

1,8

01

(18.2

6)

1–40

27.8

16.6

028

113.0

66

(p<

.001)

35–44

2,4

31

(24.6

5)

0–40

29.8

66.4

931

45–54

3,0

10

(30.5

2)

0–40

29.5

76.8

130

55–64

1,9

22

(19.4

8)

0–40

30.4

57.2

532

65þ

700

(7.1

0)

1–40

32.7

86.8

835

Stat

e(N¼

9,9

43)

AC

T174

(1.7

5)

10–40

28.9

57.2

530

41.4

59

(p<

.001)

NSW

2,7

77

(27.9

3)

0–40

29.2

57.1

630

NT

135

(1.3

6)

2–40

29.6

77.1

131

QLD

1,7

87

(17.9

7)

0–40

30.1

76.7

931

SA817

(8.2

2)

0–40

30.1

76.4

431

TA

S307

(3.0

9)

3–40

28.9

37.0

030

(con

tinue

d)

63



Tab

le3

(co

nti

nu

ed

)

Tota

lJo

bSa

tisf

action

Score

sK

rusk

al-W

allis

Tes

tPro

file

Cat

egory

n(%

)R

ange

MSD

Mdn

X2

(Sig

nifi

cance

)

VIC

3,0

02

(30.1

9)

0–40

29.6

46.7

730

WA

944

(9.4

9)

0–40

30.4

06.8

831

Rura

lity

(N¼

9,9

43)

Maj

or

city

7,6

77

(77.2

1)

0–40

29.6

26.9

231

14.4

58

(p¼

.006)

Inner

regi

onal

1,5

15

(15.2

4)

0–40

29.8

46.8

631

Oute

rre

gional

511

(5.1

4)

0–40

30.5

06.8

831

Rem

ote

194

(1.9

5)

0–40

29.5

96.7

531

Ver

yre

mote

46

(0.4

6)

19–40

31.6

15.5

732.5

Hours

work

ed(N¼

9,6

73)

<30

hr

1,5

21

(15.7

2)

0–40

31.9

76.3

333

434.6

(p<

.001)

30–49.5

hr

4,5

42

(46.9

6)

0–40

30.2

96.5

631

50þ

3,6

10

(37.3

2)

0–40

28.1

07.1

429

64



satisfaction, doctors working less than 30 hr per week had the highest

median job satisfaction score (33), compared to doctors working 30–49.5 hr

per week (31) and doctors working 50 hr or more per week (29). This held

true for all doctor types in relation to mean and median job satisfaction

except for specialists in training, where the median total job satisfaction

increased slightly from the less than 30 hr category to the 30–49.5 hr category,

even though the mean score decreased.

Discussion

The primary aim of this study was to assess the reliability and validity of a

10-item job satisfaction scale, which has been widely employed in medical

practitioner research over the last two decades and recently employed in the

MABEL survey of the Australian clinical medical workforce but for which

no validation studies have been reported. Rose (2001) has argued that, as a

consequence of the routine failure to adhere to the fundamentals of mea-

surement and examine the validity of job satisfaction instruments, there has

been insufficient recognition of the complexity of job satisfaction in

research. Not surprisingly, perhaps, there is limited agreement about which

attributes of job satisfaction need to be taken into account in its measure-

ment (van Saane, Sluiter, Verbeek, & Frings-Dresen, 2003; Weiss, 2002).

Nonetheless, there is substantial support for ensuring that job satisfaction

instruments focus on the measurement of evaluative judgments of both

extrinsic and intrinsic rewards (Rose, 2003; Spector, 1997; Warr et al.,

1979). In the present study, however, there was no support found for the dif-

ferentiation of intrinsic and extrinsic rewards as separate factors in the mea-

surement of job satisfaction.

The 10-item job satisfaction scale employed in the MABEL survey

reflects key intrinsic and extrinsic facets of job satisfaction of relevance

to doctors engaged in clinical practice across the private and public sectors

in a brief and easy to use format. This single-factor scale includes attributes

accounting for key domains contributing to total job satisfaction identified

in the broader literature (Spector, 1997; van Saane et al., 2003). Further-

more, the number of response items in the scale is substantially fewer than

other prominent, validated multiple factor scales employed with general

occupation groups such as the Job Satisfaction Survey with 36 response

items (Spector, 1997), the Job Descriptive Index with 72 response items

(Balzer et al., 1997; Spector, 1997), and the Minnesota Satisfaction Ques-

tionnaire with 100 response items and 20 in a short form (Weiss, Dawis,

England, & Lofquist, 1967). It also compares favorably with other

Hills et al. 65



multiple-factor job satisfaction instruments that have been employed

specifically in medical and health settings (Krogstad, Hofoss, Veenstra,

& Hjortdahl, 2006; Williams et al., 1999).

Factor analytic and internal reliability studies conducted on over 9,900

responses to the 10-item job satisfaction scale, from the baseline wave of

the annual MABEL survey, support the application of the instrument as a

single-factor scale. Consequently, the outcomes of this research provide a

sound basis for the determination of a composite scale score as an indica-

tive, numerical measure of job satisfaction for individuals and subpopula-

tions engaged in clinical medical practice in Australia. This will

ultimately enable the determination of the strength of relationships between

job satisfaction and a range of personal profile, health, well-being and per-

formance variables, and its role in mediating workforce participation out-

comes, including decisions to reduce working hours, change practice

location, or leave clinical practice.

The results of this study provide support for both the application of the

scale in previously conducted research and for the future applications of the

scale in medical practitioner populations. There is also potential for broader

use of the scale among medical practitioners, particularly in population-based

studies. The inclusion of other profession-specific or job-specific attributes in

the scale and undertaking subsequent factor analytic studies may also be war-

ranted if the scale was to be applied to solely salaried medical practitioners or

other health professionals, such as nurses, psychologists, or social workers,

where concerns such as promotion, workplace participation, and industrial

relations may be more relevant. Importantly, where more in-depth, diagnostic

approaches to job satisfaction are required, such as in circumstances where

job satisfaction is determined to be problematic, the employment of a more

inclusive, multifactor instrument would need to be considered.

The secondary aim of this study was to demonstrate the application of

the scale in the MABEL cohort. In developing a composite score for job

satisfaction in Australian doctors, important information about the similari-

ties and differences of job satisfaction in Australian medical clinicians can

be determined. Job satisfaction outcomes from the MABEL survey accord

with results reported in much of the recent international research literature,

indicating that medical practitioners experience high levels of job satisfac-

tion overall (Aasland et al., 2010; Davenport et al., 2008; Kumar, Fischer,

Robinson, Hatcher, & Bhagat, 2007; Lindfors et al., 2007; Linzer et al.,

2009; McNearney et al., 2008; Rosta et al., 2010), despite earlier concerns

about plummeting satisfaction in medicine (Edwards, Kornacki, & Silver-

sin, 2002; McGlone & Chenoweth, 2001; Weinstein & Wolfe, 2007).




In relation to gender, no significant differences were detected in total job

satisfaction scores between male and female medical practitioners in the

MABEL survey, which reflects the general findings of numerous studies

(Emmons, Nichols, Schulkin, James, & Cain, 2006; Kumar et al., 2007;

Lindfors et al., 2007; McNearney et al., 2008; Rosta et al., 2009; Walker

& Pirotta, 2007). Rural practice location was also identified as being signif-

icantly associated with higher levels of job satisfaction in Australian doc-

tors, as determined in previous research (Harris et al., 2007; Ulmer &

Harris, 2002), but job satisfaction appears to be highest for Australian doc-

tors in remote locations.

Reflecting the results of other medical practitioner job satisfaction

research, younger doctors were found to be less satisfied with their work

(McNearney et al., 2008; Pathman et al., 2002; Rosta et al., 2009; Whalley

et al., 2006), as were doctors working longer hours (French et al., 2004;

Katerndahl et al., 2009; Simoens et al., 2002; Ulmer & Harris, 2002).

A potential ‘‘dose–response’’ relationship between hours worked per week

and total job satisfaction was detected. The demonstration of a clear dose–

response relationship between variables can provide strong supportive

evidence of a causal relationship (Bonita, Beaglehole, & Kjellstrom,

2006), but other factors may have contributed to this finding. The deter-

mination that hospital nonspecialists and specialists in training experience

significantly lower levels of job satisfaction is a more novel outcome. This

may not be unexpected, however, since hospital non-specialists and

specialists-in training are younger and work longer hours (Australian

Institute of Health and Welfare, 2009; Joyce et al., 2010). Such an out-

come suggests that extra supports, including relief from long hours of

work, may be required for these doctor groups, particularly younger prac-

titioners, to ensure that they are not lost from the profession.

Job satisfaction in medicine reflects a range of interacting monetary,

personal, work, and patient-related factors affecting the working lives of

clinicians. Consequently, the application of the 10-item job satisfaction

scale has considerable potential to provide indicative data about the quality

of doctors’ working lives, their commitment to clinical practice, and the

likelihood of communities and populations being able to access high-

quality medical care into the future. There is considerable evidence in the

literature that appropriate income and resources, adequate time, reasonable

workloads, perceived work autonomy and control, and co-operative work-

ing relationships impact positively on job satisfaction in medicine. It could

be expected, therefore, that an acceptable balance of these factors with the

need to manage the increasingly complex care needs of patients, in often

Hills et al. 67



demanding clinical environments, will be required to maintain clinicians’

satisfaction with their work. Since job satisfaction is associated with the

quality of medical care, and has been shown to be predictive of intentions

to leave patient care or leave medicine altogether, such concerns are likely

to be of even greater import where there are existing challenges in the

recruitment and retention of medical clinicians.

There are some limitations to this study. First, a self-selected cohort of

Australian doctors provided data in a self-report survey. While this places

some constraints on generalizing the findings to the total population of

Australian doctors, the overall representativeness of the MABEL survey data

has been established (Joyce et al., 2010). Second, cross-sectional data from

the first wave of the annual MABEL survey were utilized and only limited

statistical analyses of associations between selected variables was conducted.

Consequently, more complex interrelationships between variables and caus-

ality were not determined. Third, the application of the 10-item job satisfac-

tion scale in the MABEL cohort provided some supportive evidence of the

validity and discriminatory properties of the scale. The relative brevity of the

scale, however, may have resulted in the exclusion of important facets of job

satisfaction in clinical medicine, affecting the capacity of the scale to

accurately estimate and discriminate between different levels of job satis-

faction in clinical medical practice. Lastly, while the factor analytic pro-

cedures in this study have been applied to the responses of a large cohort

of medical practitioners and replicated in four subpopulations of the

cohort, the determination of a single-factor solution may not be general-

izable to other medical practitioner and health professional populations.

This may, for example, be due to variations in a range of endogenous and

exogenous factors not only relating to specific aspects of professional

practice but also to a range of individual, work, familial, community, and

country-specific characteristics and conditions.

Conclusions

The validation of a widely used 10-item job satisfaction scale in a large

cohort of Australian medical practitioners supports the employment of the

scale in ongoing exploratory and comparative research in medical practi-

tioner populations internationally. Researchers should find a level of confi-

dence in utilizing the scale to determine a composite score for job

satisfaction. This could enable the establishment of comparisons or relation-

ships with other key variables in study populations or population samples

and facilitate the identification and prediction of patterns of characteristics




or performance in populations of interest. Within the context of the MABEL

longitudinal survey, some patterns of job satisfaction among Australian

doctors have been determined, including the indication of a possible causal

association between increasing hours of work and decreasing job satisfac-

tion. The validation of the 10-item job satisfaction scale is a fundamental

step in ongoing, longitudinal research that is expected to provide important

insights into the interrelationships between job satisfaction and a broad

range of variables that may affect and predict the patterns and determinants

of clinical medical workforce participation in Australia, into the future.

Declaration of Conflicting Interests

The authors declared no potential conflicts of interests with respect to the

authorship and/or publication of this article.

Funding

The authors disclosed receipt of the following financial support for the research

and/or authorship of this article: The Medicine in Australia: Balancing Employ-

ment and Life study was supported by a National Health and Medical Research

Council Health Services Research Grant (454799) and the Commonwealth

Department of Health and Ageing in Australia.

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