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RESEARCH ARTICLE
Revisiting the factor structure of the Short-
Form McGill Pain Questionnaire-2 (SF-MPQ-2):
Evidence for a bifactor model in individuals
with Chiari malformation
David M. Tokar
1
*, Kevin P. Kaut
2
, Philip A. AllenID
2
1Department of Social and Behavioral Sciences, Central State University, Wilberforce, OH, United States of
America, 2Department of Psychology, University of Akron, Akron, OH, United States of America
*dtokar@centralstate.edu
Abstract
The Short-Form McGill Pain Questionnaire-2 (SF-MPQ-2; Dworkin et al., 2009) is intended
to measure the multidimensional qualities of pain (i.e., continuous, intermittent, neuropathic,
and affective) as well as total pain. Using structural equation modeling, we evaluated the fit
of four competing measurement models of the SF-MPQ-2—an oblique 4-factor model, a 1-
factor model, a higher-order model, and a bifactor model—in 552 adults diagnosed with
Chiari malformation, a chronic health condition whose primary symptoms include head and
neck pain. Results revealed the strongest support for the bifactor model, suggesting that
SF-MPQ-2 item responses are due to both a general pain factor and a specific pain factor
that is orthogonal to the general pain factor. Additional bifactor analyses of the SF-MPQ-2’s
model-based reliability and dimensionality revealed that most of the SF-MPQ-2’s reliable
variance is explained by a general pain factor, and that the instrument can be modeled unidi-
mensionally and scored as a general pain measure. Results also indicated that the general
and affective pain factors in the bifactor model uniquely predicted pain-related external crite-
ria (e.g., depression, anxiety, and stress); however, the continuous, intermittent, and neuro-
pathic factors did not.
Introduction
In the 2019 National Health Interview Survey, 50.2 million adults (20.5%) claimed that they
experienced pain on most or every day [1–3], and this estimate will undoubtedly increase as
the population continues to age [3]. Given the tremendous impact of chronic pain on individ-
uals and society, it is not surprising that chronic pain has become a topic of increasing interest
to psychologists [4]. Since the 1965 publication of Melzack and Wall’s seminal gate control the-
ory of pain, [5] considerable research has supported the notion that the chronic pain experi-
ence is subjective and multidimensional, consisting of biological, psychosocial, and behavioral
components [4,6–8].
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Citation: Tokar DM, Kaut KP, Allen PA (2023)
Revisiting the factor structure of the Short-Form
McGill Pain Questionnaire-2 (SF-MPQ-2): Evidence
for a bifactor model in individuals with Chiari
malformation. PLoS ONE 18(10): e0287208.
https://doi.org/10.1371/journal.pone.0287208
Editor: Leica S. Claydon-Mueller, Anglia Ruskin
University UK, UNITED KINGDOM
Received: October 28, 2022
Accepted: June 1, 2023
Published: October 5, 2023
Peer Review History: PLOS recognizes the
benefits of transparency in the peer review
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all of the content of peer review and author
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editorial history of this article is available here:
https://doi.org/10.1371/journal.pone.0287208
Copyright: ©2023 Tokar et al. This is an open
access article distributed under the terms of the
Creative Commons Attribution License, which
permits unrestricted use, distribution, and
reproduction in any medium, provided the original
author and source are credited.
Data Availability Statement: All relevant data are
within the paper and its Supporting Information
files.
Clearly, reliable and valid assessment tools are required for the systematic investigation,
accurate classification, and effective treatment of chronic pain conditions [6,7]. Over the past
40 years, the McGill Pain Questionnaire (MPQ [9]) and the Short-Form McGill Pain Ques-
tionnaire (SF-MPQ [10]) have been among the most widely used self-report measures of pain
qualities. Research has tended to support the reliability and validity of the different versions of
the MPQ, including the most recently revised and expanded version of the SF-MPQ
(SF-MPQ-2; Dworkin et al., 2009 [11]); however, the evidence regarding the factor structure of
the SF-MPQ-2 is equivocal.
The organization of the SF-MPQ-2 underscores the multidimensional perspective of pain,
with items clustered into four subscales reflecting the subjectively experienced qualities of pain
as continuous (e.g., “throbbing pain”; “cramping pain”; “gnawing pain”), intermittent (e.g.,
“shooting pain”; “sharp pain”; “stabbing pain”), affective (e.g., “tiring-exhausting”; “punish-
ing”; “fearful”), and neuropathic (e.g., “hot-burning pain”; “cold-freezing pain”; “itching
pain”). This measure provides users with an efficient assessment of these pain dimensions, as
well as total pain. As such, it is likely to be the instrument of choice in future clinical research.
The primary purpose of the current study was to evaluate further the factor structure of the
SF-MPQ-2 in a large sample of individuals diagnosed with a chronic health condition whose
primary symptoms manifest as head and neck pain (i.e., Chiari malformation; see Method for
clinical description).
It should be noted that Dworkin et al. (2009) [11] developed the SF-MPQ-2 in order to
assess the growing chronic health concerns associated with ‘neuropathic pain’. To some extent,
the emergence of neuropathic pain as a significant and chronic health issue reflects evolving
societal health concerns and outcomes, coupled with advances in pain research and diagnostic
specificity [12]. Causes of neuropathic pain (i.e., defined as a disease or lesion of the somato-
sensory system) likely reflect, in part, sequelae associated with an aging population, including
the incidence of diabetes, neurodegenerative conditions (e.g., Parkinson’s disease), and stroke;
in addition, diverse nervous system pathologies such as post-surgical pain, spinal cord injury,
and multiple sclerosis are included here, as well as the effects of other diseases such as cancer
and HIV infection (see Colloca et al., 2017, for comprehensive review) [12].
Dworkin et al. [11] specifically addressed the issue of neuropathic pain in their develop-
ment of the SF-MPQ-2 by adding nine items to the original 15 of the SF-MPQ. Using a web-
based sample of 882 chronic pain patients (i.e., neuropathic pain, n= 349, non-neuropathic
pain, n= 533), exploratory factor analyses (EFAs) on 20 sensory pain items (i.e., four affective
pain items were not included, but were retained for the final version of the SF-MPQ-2)
resulted in a three-factor structure that was generally invariant across the neuropathic pain
and non-neuropathic pain groups. After dropping two items loading only moderately on the
second factor, EFAs on the remaining 18 sensory pain items again yielded a three-factor struc-
ture that was generally invariant across the neuropathic pain and non-neuropathic pain
groups. Based on their EFA results and prior research on symptoms of neuropathic and non-
neuropathic pain, the four subscales comprising the SF-MPQ-2 were established (i.e., continu-
ous, intermittent, neuropathic, and affective).
Psychometric properties of the SF-MPQ-2 have been respectable, with internal consistency
reliabilities (αs) ranging from .73 to .87 for the four pain subscale scores and αs of .91 and .95
for total pain scores in two separate samples [11]. Construct validity has been documented via
positive correlations with measures of pain intensity and impact. In addition, Dworkin et al.
(2009) [11] utilized confirmatory factor analysis (CFA) to evaluate the fit of their web-based
survey data to a four-factor structure consistent with the scoring of the SF-MPQ-2’s four pain
subscales. However, and of particular relevance to the present study, rather than simulta-
neously modeling all four pain factors and allowing them to covary, Dworkin et al. (2009) [11]
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Funding: This research was supported by a
Conquer Chiari Research Grant from the Conquer
Chiari Foundation as well as NIH grant
1R15NS109957-01A1. The funders did not play
any role in the study design, data collection and
analysis, decision to publish, or preparation of this
manuscript.
Competing interests: No–we have no competing
interests.
performed separate CFAs on the items composing each of the four respective SF-MPQ-2 sub-
scales. This forced restriction of covariance among items across subscales is potentially prob-
lematic, reiterating the question of whether a four-factor model adequately fits the SF-MPQ-2
data, or a more parsimonious (e.g., one-factor, higher-order) model better represents the vari-
ability in SF-MPQ-2 item responses.
Subsequent psychometric investigations of the SF-MPQ-2 have been illustrative. Using
patient samples with diverse pain conditions has yielded consistent support for the internal
consistency and convergent validity of the specific pain subscales as well as total pain scale
scores [13–15]. However, independent tests of the SF-MPQ-2’s factor structure have yielded
equivocal findings. Gauthier et al. (2014) [14], using a combined sample of 190 older and
younger patients with cancer pain, tested an oblique four-factor model corresponding to the
four intended SF-MPQ-2 pain constructs and found a “reasonable fit” (p. 763) that was shown
to be configurally invariant across the two age groups. Yet, several CFA fit index values (e.g.,
CFI = .78, TLI = .75) indicated a poor fit to the data. Moreover, latent factor correlations were
high (median rs of .71 and .85 in the younger and older groups, respectively), suggesting that a
higher-order structure, in which the first-order factors load onto a superordinate general pain
factor, might provide a better fit. Dworkin et al. (2015) used CFA to test the fit of a four-factor
model corresponding to the SF-MPQ-2’s four intended pain constructs to data obtained from
666 acute lower back pain patients. Results indicated a good fit for the Continuous subscale
data, but mixed or no support for the Intermittent, Neuropathic, and Affective subscale data.
Like Dworkin et al. (2009) [11], Dworkin et al. (2015) [13] performed separate CFAs on each
set of items corresponding to a specific SF-MPQ-2 pain construct instead of testing the
dimensionality of the total measure. As such, conclusions regarding the structure of the
SF-MPQ-2 remain equivocal, and certainly underscore the need for continued efforts to assess
the psychometric properties of the scale.
In perhaps the most comprehensive examination of the SF-MPQ-2’s factor structure, Love-
joy et al. (2012) [15] evaluated the fit to their data (N= 186 veterans with chronic pain symp-
toms) of three different structural models. The first model was an oblique four-factor model
corresponding to the SF-MPQ-2’s four intended pain constructs. The second model was a
one-factor model corresponding to a total pain construct in which all 22 SF-MPQ-2 items
loaded onto one general factor. The third model was a higher-order model in which the
SF-MPQ-2 items loaded onto one of four domain-specific pain factors, which in turn contrib-
uted to a higher-order general pain factor. CFA results revealed the strongest support for the
oblique four-factor and higher-order models. Support for the oblique four-factor model
implies four related but distinct pain constructs. Support for the higher-order model implies a
hierarchical structure in which a general pain factor representing the shared variance among
the first-order pain factors ultimately (i.e., and indirectly via the first-order factors) accounts
for variation in SF-MPQ-2 item responses. That is, support for the higher-order model implies
that the SF-MPQ-2 item responses contain variance attributable to specific pain constructs
(e.g., continuous, intermittent, neuropathic, affective) as well as a higher-order general pain
construct [16].
Although Lovejoy et al.’s (2012) [15] CFA results suggest the possibility that an oblique
four-factor model or a higher-order model best represents the structure of the SF-MPQ-2,
their findings must be interpreted cautiously for at least two reasons. First, neither the oblique
four-factor model nor the higher-order model met recommended fit index value cutoffs for
adequate fit (e.g., CFI close to .95, RMSEA .06; Kline, 2016) [17]. Second, direct (e.g., χ
2
dif-
ference) tests comparing the fit of the different nested models were not performed; therefore,
the relative superiority of the oblique four-factor and higher-order models is questionable.
Thus, a primary purpose of the present study is to evaluate the absolute and relative fit of an
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oblique four-factor model, a one-factor general model, and a higher-order model (see Fig 1)
for SF-MPQ-2 data in a large sample (N= 552) of patients diagnosed with Chiari malforma-
tion, a condition associated with skeletal abnormalities (often congenital in nature), and com-
monly presenting with chronic pain and related health complications (see the Participants
section below for a more detailed description of Chiari malformation).
In our work, we also considered an alternative bifactor model, which has yet to be evaluated
with the SF-MPQ-2. In contrast to the higher-order model, bifactor models assume two inde-
pendent sources of common variance directly influencing all of the SF-MPQ-2 item responses
(see Fig 1). That is, bifactor models imply that item responses are potentially influenced by
both a general factor (e.g., overall pain) and a specific factor (i.e., one of the four pain dimen-
sions of the SF-MPQ-2) that is orthogonal to the general factor and the other specific factors.
Estimates from bifactor models enable researchers to determine whether domain-specific fac-
tors explain variability in item responses independent of a general latent factor. Thus, bifactor
models can be used to help clarify the dimensionality of a measure and determine whether
scoring an instrument for raw total and raw subscale scores is warranted [18]. Assuming a
bifactor model best represents the structure of the SF-MPQ-2, additional bifactor analyses
(e.g., model-based reliability, explained common variance; see Rodriguez et al., 2016a, 2016b)
Fig 1. Higher-order model depicted on the top. Bifactor model depicted on the bottom. SF-MPQ-2 item numbers are presented in the rectangles.
https://doi.org/10.1371/journal.pone.0287208.g001
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[18,19] could shed light on the appropriateness of scoring and interpreting the instrument for
four domain-specific pain constructs [11,14,15] and for a general total pain construct [11,15].
Finally, assuming the SF-MPQ-2 structure is best represented by a bifactor model, we inves-
tigated the incremental validity of the SF-MPQ-2 general and specific factors (e.g., continuous
pain) by examining their unique relations with several external criteria (i.e., symptom severity,
neck pain, depression, anxiety, and stress). Consistent with previous research demonstrating
moderate to strong positive associations of sensory (i.e., continuous, intermittent, and neuro-
pathic), affective, and total pain scores with pain severity [15] and intensity [13,14], we hypoth-
esize that continuous, intermittent, neuropathic, affective, and general pain factor scores will
relate uniquely and positively with symptom severity and neck pain. Based on previously
reported positive associations of anxiety and depression with all four types of (and total) pain
[13,15], we hypothesize that continuous, intermittent, neuropathic, affective, and general pain
factor scores will relate uniquely and positively with anxiety, depression, and stress. Because
the construct reliability of the general pain factor is likely to be higher than that of the “residua-
lized” specific pain factors (Rodriguez et al., 2016b, p. 146) [18], we anticipate the strongest
unique associations between the general pain factor and the external criteria.
Method
Ethics statement
The research reported in the present paper was approved by the Institutional Review Board at
The University of Akron. All participants provided electronic informed consent.
Participants and procedure
We collected data using a web-based survey containing the SF-MPQ-2 and other measures
described below. We recruited a sample of 552 adult participants (i.e., 18 years of age) from
the Chiari 1000 database (i.e., a database established for the collection of behavioral and ana-
tomical information from Chiari malformation patients; see Allen et al., 2018) [20] who had
been diagnosed with Chiari malformation (CM). This malformation is often congenital in
nature, impacting cranial and spinal column anatomy, and typically involving the base of the
skull, upper vertebral column, and associated neural tissue [21]. In particular, the radiological
definition of CM specifies this anatomical issue, involving portions of the cerebellum (i.e., cer-
ebellar tonsils) extending at least 5 mm below the line of demarcation between the brain and
the spinal cord—an opening in the base of the skull known as the foreman magnum [22]. Etio-
logical, diagnostic, and treatment considerations continue to emerge, particularly in the con-
text of modern neuroimaging advances [21,23,24]. Not uncommonly, reduced cerebrospinal
fluid volume in the posterior portions of the skull (i.e., posterior cranial fossa, beneath the
occipital lobes) and the anterior cerebrospinal fluid (CSF) space in the upper cervical region of
the spinal canal exacerbate cervico-medullary compression by the cerebellum, thereby exerting
pressure (i.e., particularly during diastolic cardiac cycles) on underlying brainstem and cervical
spine regions [21,25,26]; see also [24]. Also, pain levels (as assessed by the SF-MPQ-2) in CM
relative to healthy age- and education-matched controls are associated with significantly differ-
ent levels of white-matter integrity as indexed by diffusion tensor imaging (DTI, Houston
et al., 2020 [27]) as well as intrinsic functional connectivity as indexed via resting-state func-
tional magnetic resonance imaging (fMRI) (Houston et al., 2021 [28]).
Variations in the degree of structural abnormality naturally influence functional and sen-
sory outcomes, although the most commonly experienced physical symptoms include chronic
pain, headache, and sensory-motor disturbances typically affecting regions of the head, neck,
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and upper extremities [21]. Increased levels of anxiety and depression are also observed in CM
[29] (Garcia et al., 2019), as well as cognitive dysfunction [29–33].
In the present sample, the most frequently reported comorbid illnesses were autoimmune
disease (40.22%), scoliosis (23.95%), syringomyelia (22.85%), pseudotumors (10.24%), and
Ehlers-Danlos syndromes (9.69%). Demographic data were based on 548 (99.3% of the total
sample) cases for whom such data were available. Participants were 518 (94.5%) women and
30 (5.5%) men who ranged in age from 18 to 66 years (M= 37.80, SD = 10.92). The majority
(88.1%) of participants identified ethnically as White/European American, with 4.6% Black/
African American, 4.2% Hispanic, 2.4% American Indian/Native American, 0.5% Asian/Asian
American, and 0.2% Pacific Islander. In terms of employment status, 53.5% indicated that they
were not currently working, whereas 46.5% indicated that they were currently working.
Regarding highest level of education completed, 27.0% indicated some college, 20.8% high
school graduate, 15.3% bachelor’s degree, 13.5% associate’s degree, 10.9% trade school/techni-
cal school, 10.9% master’s degree, and 1.5% doctorate degree.
All participants scored a minimum of 1 or above (i.e., indicating at least some degree of
pain intensity) on one or more of the four SF-MPQ-2 pain subscales (i.e., continuous, inter-
mittent, neuropathic, or affective). Chiari malformation is not considered to be a neuropathic
disorder (i.e., involving disease or lesion of the somatosensory system; see Treede et al., 2008
[34]); therefore, we assume that our sample is primarily non-neuropathic in nature. Institu-
tional Review Board approval was obtained before the onset of this study, and all participants
consented before they participated.
Measures
In addition to the SF-MPQ-2, we collected data on the Neck Disability Index (Vernon & Mior,
1991 [35]), self-rated Chiari symptom severity, and the Depression Anxiety Stress Scales-21
(DASS-21; Lovibond & Lovibond, 1995 [36]).
SF-MPQ-2
The SF-MPQ-2 [11] is a 22-item self-report measure of different pain qualities or related
symptoms. Participants used an 11-point Likert scale (0 = none, 10 = worst possible) to indicate
the intensity of each pain quality/symptom experienced within the past week. The SF-MPQ-2
is scored for continuous (6 items; e.g., “throbbing”), intermittent (6 items; e.g., “shooting”),
neuropathic (6 items; e.g., “tingling”), and affective (4 items; e.g., “tiring-exhausting”) pain as
well as total pain by calculating the mean for each subscale (or total), with higher scores corre-
sponding to more intense pain symptoms. Dworkin et al. (2009) reported Cronbach’s alphas
of .91 (total pain; current α= .94), .73 (Continuous pain; current α= .82), .85 (Intermittent
pain; current α= .89), .78 (Neuropathic pain; current α= .83) and .77 (Affective pain; current
α= .82) in a sample of 882 adults with chronic pain. Gauthier et al. (2015) [14] demonstrated
support for the convergent validity of the SF-MPQ-2 through expected relations with measures
of pain intensity, interference, and relief; depressive symptoms; and physical and mental health
quality of life.
Neck Disability Index. We used the Neck Disability Index [35] (NDI), an adapted version
of the Oswestry Low Back Pain Disability Questionnaire (Fairbanks, Couper, Davies, &
O’Brien, 1980) [37], to assess neck pain disability, a common symptom of Chiari malformation
[38]. Participants used a 6-point Likert scale (0 = no interference [e.g., “I can do as much work
as I want to”], 5 = extreme interference [e.g., “I cannot do any work at all”]) to indicate the
extent to which their neck pain has affected their ability to perform eight different everyday
activities (e.g., working, driving), as well as their experience of pain intensity (0 = “I have no
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pain at the moment,” 5 = “The pain is the worst imaginable at the moment”) and headaches (0
= “I have no headaches at all,” 5 = “I have headaches almost all the time”). Scores for each of
the 10 items are summed and then doubled; thus, total scores can range from 0–100, with
higher scores indicating greater disability. McCarthy, Grevitt, Silcocks, and Hobbs (2007) [39]
reported a Cronbach’s alpha of .86 (current α= .89) and demonstrated support for the concur-
rent validity of the NDI through expected relations with the Short Form 36 Health Survey
Questionnaire (SF36; Brazier et al., 1992) [40], a measure of functional ability and overall
health and well-being, in a sample of 160 patients with neck pain.
Symptom severity. Participants used a 4-point Likert-scale (1 = mild, 4 = very severe) to indi-
cate their overall level of Chiari malformation symptom severity, which is characterized by chronic
headache and neck pain [38], anxiety and depression [29], and cognitive dysfunction [30–33].
Depression Anxiety Stress Scales-21. We used the 21-item Depression Anxiety Stress
Scales-21 (DASS-21) [36] to measure self-reported levels of depression, anxiety, and stress.
Each of the three DASS-21 subscales is composed of seven items. Participants used a 4-point
Likert scale (0 = did not apply to me at all; 3 = applied to me very much,or most of the time) to
indicate how frequently they experienced each within the past week. Sample items include “I
couldn’t seem to experience any positive feeling at all” (Depression), “I felt scared without any
good reason” (Anxiety), and “I found it hard to wind down” (Stress). Scores for each DASS-21
subscale are summed and multiplied by two (for comparability with the DASS-42), with higher
scores corresponding to higher levels of each construct. Page, Hooke, and Morrison (2007)
[41] reported Cronbach’s alphas of .96 (Depression; current α= .92), .92 (Anxiety; current α=
.83), and .95 (Stress; current α= .85) in a sample of 124 adult patients diagnosed with depres-
sion. Henry and Crawford (2005) demonstrated support for the convergent validity of the
DASS-21 via expected relations with independent measures of depression and anxiety.
Results
Missing data
Seventy-three of the 552 cases contained at least one missing item-level data point, and the
overall rate of missing data was 6.18%. Little’s (1988) [42] missing completely at random
(MCAR) test was nonsignificant, χ
2
(67) = 52.68, p= .90, indicating that missingness was not
systematically related to any of the study variables. We used full information maximum likeli-
hood (FIML) estimation to deal with missing data in all structural equation modeling (SEM)
analyses. FIML uses all available data to generate unbiased parameter estimates and standard
errors (Tabachnick & Fidell, 2013) [43].
Descriptive statistics and intercorrelations
Prior to conducting structural equation modeling (SEM) analyses, all measured variables were
evaluated for normality and outliers. All variables satisfied assumptions of univariate normal-
ity (i.e., absolute skewness and kurtosis values 2; Tabachnick & Fidell, 2013) [43], and no
univariate outliers were identified. However, Mardia’s (1970) [44] tests for multivariate skew-
ness and kurtosis indicated violations of multivariate normality; therefore, all SEM analyses
were performed using a robust estimator (i.e., MLR) to correct for multivariate non-normality
(Muthe
´n & Muthe
´n, 1998–2017) [45].
Means, standard deviations, alpha reliabilities, and intercorrelations of all measured vari-
ables are reported in Table 1. SF-MPQ-2 total and subscale score means were comparable (i.e.,
within roughly half a standard deviation) to those reported by Dworkin et al. (2009) [11] based
on samples of chronic pain patients. SF-MPQ-2 alphas (ranging from .82 to .89 for subscales,
.94 for total pain scores) were comparable to those reported in other chronic pain patient
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samples (Dworkin et al., 2009 [11]; Lovejoy et al., 2012 [15]). SF-MPQ-2 subscale intercorrela-
tions were large (median r= .72), and correlations between total and subscale scores (mdn =
.89) approached unity, suggesting that the subscales may not be measuring unique pain con-
structs. Associations of SF-MPQ-2 total and subscale scores with measures of neck pain, symp-
tom severity, depression, anxiety, and stress were positive and (except for those involving
symptom severity) of moderate to large magnitude.
SP-MPQ-2 factor structure
We performed SEM analyses using Mplus version 8.1 (Muthe
´n & Muthe
´n, 1998–2018) [45] to
examine the fit of four different structural models (described below) to our data. Model fit was
evaluated using the Satorra-Bentler scaled (i.e., mean-adjusted) χ
2
goodness-of-fit test, com-
parative fit index (CFI), root mean square error of approximation (RMSEA), and standardized
root mean square residual (SRMR). CFI values close to .95, RMSEA values .06, and SRMR
values .08 indicate a good fit (Kline, 2016) [17]. We used the Satotta-Bentler scaled χ
2
differ-
ence test to compare the goodness-of-fit of the different nested models and differences in the
other fit index values to compare nonnested models.
Model 1: Oblique four-factor model. The first model was an oblique four-factor model
based on Dworkin et al.’s (2009) recommended scoring of the SF-MPQ-2 [11]. Continuous,
intermittent, neuropathic, and affective pain constructs were modeled as correlated latent fac-
tors. Items scored for each SF-MPQ-2 subscale were indicators of the corresponding latent fac-
tor. As shown in Table 2, fit index values indicated that Model 1 did not fit the data well.
Model 2: One-factor model. The second model was a one-factor general pain model with
all 22 SF-MPQ-2 items as indicators of a single latent factor. As shown in Table 2, Model 2 had
a poor fit to the data. Relative to Model 1, Model 2 provided a significantly worse fit, Δχ
2
(6,
N= 552) = 178.35, p<.001.
Model 3: Higher-order model. The third model was a higher-order model with four first-
order pain factors identical to those in Model 1 as well as a higher-order general pain factor
onto which the first-order factors loaded. As shown in Table 2, Model 3 did not fit the data
well. However, Model 3 was not a significantly worse fit than Model 1, Δχ
2
(2, N= 552) =
5.94, p>.05.
Table 1. Intercorrelations, internal consistencies, means, and standard deviations of all variables.
Variable 1 2 3 4 5 6 7 8 9 10
1. Total Pain –
2. Continuous Pain .90 –
3. Intermittent Pain .91 .74 –
4. Neuropathic Pain .88 .70 .73 –
5. Affective Pain .85 .73 .69 .64 –
6. Neck Pain .64 .58 .61 .52 .53 –
7. Symptom Severity .20 .16 .21 .13 .19 .26 –
8. Depression .38 .31 .35 .28 .43 .42 .15 –
9. Anxiety .48 .42 .40 .41 .50 .48 .16 .68 –
10. Stress .39 .33 .33 .30 .45 .34 .09 .68 .73 –
α.94 .82 .89 .83 .82 .89 NA .92 .83 .85
M4.16 4.61 3.95 3.82 4.33 48.05 2.73 13.31 13.16 17.37
SD 2.26 2.38 2.75 2.43 2.71 18.41 0.85 11.15 10.14 10.13
Note.Ns ranged from 479–552. All correlations are significant at p<.05.
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Model 4: Bifactor model. The fourth model was a bifactor model in which the indicator
variables (i.e., items) had loadings on both a specific pain factor corresponding to one of the
four factors in Model 1 and a general pain factor. All factor covariances were fixed to zero.
Although the χ
2
value was statistically significant (indicating a non-perfect fit), the other fit
index values indicated that Model 4 provided a good fit to the data (see Table 2). Furthermore,
Model 4 provided a significantly better fit than did the more parsimonious higher-order
model (Model 3), Δχ
2
(18, N= 552) = 479.21 p<.001. Overall, results indicated that the bifac-
tor model was the best-fitting model in our data.
As shown in Table 3, all 22 items loaded significantly and substantively (ranging from .47 to
.75, mdn = .65) on the general pain factor. Conversely, only 12 of 22 items loaded significantly
on a specific pain factor, and those factor loadings varied considerably in terms of magnitude.
For example, the four significant neuropathic pain factor loadings were .11 (item 19), .14 (item
20), .60 (item 22), and .77 (item 21). Furthermore, general pain factor loadings equaled or
exceeded specific pain factor loadings for 21 of the 22 items.
Factor loadings in bold are significant at p<.05.
Model-based reliability and dimensionality
Next, we calculated a number of additional bifactor indices to evaluate the reliability of the
SF-MPQ-2 total and subscale pain scores as well as the dimensionality of the SF-MPQ-2.
Model-based estimates of internal consistency of the total and subscale scores were calculated
using coefficient omega and coefficient omega hierarchical. Omega (ω) and omega subscale
(ω
S
) reflect the proportion of variance in total scores and subscale scores, respectively,
accounted for by all common variance sources (i.e., the general factor and corresponding spe-
cific factor[s]) (Rodriguez et al., 2016a, 2016b) [18,19]. Omega hierarchical (ω
H
) reflects the
proportion of total score variance explained by the general factor after accounting for the vari-
ance attributed to the specific factors, whereas omega hierarchical subscale (ω
HS
) reflects the
proportion of subscale score variance explained by the corresponding specific factor after
removing the variance attributed to the general factor (Rodriguez et al., 2016a, 2016b) [18,19].
High ω
H
values suggest that total scores are due primarily to a general factor, whereas high ω
HS
values suggest that subscale scores are due primarily to a specific factor common to the items
composing that subscale. Although there are no clear cutoffs for evaluating ω
H
and ω
HS
, Reise,
Bonifay, and Haviland (2013) [46] recommended “that a minimum would be greater than .50,
and values closer to .75 would be much preferred” (p. 137) for determining the unique infor-
mation provided by total and subscale scores.
Omega coefficients for the total (ω= .96) and specific pain scores (ω
S
range = .84-.91) were
high (see Table 3), indicating that the total and subscale composite scores were highly reliable.
Table 2. Summary of fit statistics for all models.
Model Description χ
2
df CFI RMSEA SRMR
Model 1: Oblique four-factor 907.38 203 .87 .079 .067
Model 2: One-factor 1258.44 209 .81 .095 .057
Model 3: Higher-order 913.27 205 .87 .079 .068
Model 4: Bifactor 458.38 187 .95 .051 .038
Note.N= 552. df = degrees of freedom; CFI = comparative fit index; RMSEA = root mean square error of approximation
SRMR = standardized root mean square residual.
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The omega hierarchical coefficient (ω
H
= .91) indicated that 91% of the variance in SF-MPQ-2
total pain scores was due to differences on the general pain factor. Conversely, omega hierar-
chical subscale (ω
HS
) values, which ranged from .07-.18, indicated that only 7–18% of the vari-
ance in SF-MPQ-2 subscale scores was due to differences on the corresponding specific pain
factors. Dividing the obtained omega hierarchical (i.e., ω
H
and ω
HS
) values by corresponding
omega (i.e., ωand ω
S
) values revealed that 95% (i.e., .91/.96) of the reliable variance in
SF-MPQ-2 total pain scores was explained by the general pain factor, whereas only 9–22% of
the reliable variance in SF-MPQ-2 subscale scores was explained by the corresponding specific
pain factors. These results indicate that the vast majority of the reliable variance in SF-MPQ-2
total and subscale scores was explained by the general pain factor. Thus, results suggest that
SF-MPQ-2 total scores reliably measure their target construct (i.e., general pain); however, the
subscale scores primarily measure the general pain construct instead of their intended specific
pain constructs.
Next, we examined the explained common variance (ECV), the percentage of uncontami-
nated correlations (PUC), and the absolute relative parameter bias (ARPB) to determine
Table 3. Factor loadings for unidimensional and bifactor solutions.
Bifactor Model
SF-MPQ-2 item 1-factor Gen. Contin. Intermit. Neuro. Affect.
1 (throbbing) .64 .64 .09
5 (cramping) .59 .60 .06
6 (gnawing) .57 .59 .06
8 (aching) .64 .62 .58
9 (heavy) .71 .70 .28
10 (tender) .64 .65 .12
2 (shooting) .75 .70 .42
3 (stabbing) .77 .71 .62
4 (sharp) .77 .71 .49
11 (splitting) .75 .75 .08
16 (electric-shock) .63 .63 .07
18 (piercing) .73 .74 .08
7 (hot-burning) .65 .66 .07
17 (cold-freezing) .57 .59 .06
19 (caused by lt. touch) .66 .66 .11
20 (itching) .48 .47 .14
21 (tingling) .66 .62 .77
22 (numbness) .64 .60 .60
12 (tiring-exhausting) .62 .61 .05
13 (sickening) .72 .72 .28
14 (fearful) .63 .62 .60
15 (punishing-cruel) .66 .66 .44
ω/ω
S
.96 .84 .91 .85 .84
ω
H
/ω
HS
.91 .07 .13 .16 .18
ECV/ECV
S
.76 .04 .07 .08 .05
PUC .78
Note.N= 552. SF-MPQ-2 = Short-Form McGill Pain Questionnaire-2. Gen. = General pain factor; Contin. = Continuous pain; Intermit. = Intermittent pain; Neuro. =
Neuropathic pain; Affect. = Affective pain; ω= omega; ω
S
= omega subscale; ω
H
= omega hierarchical; ω
HS
= omega hierarchical subscale; ECV = explained common
variance; ECV
S
= explained common variance of specific factor; PUC = percentage of uncontaminated correlations.
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whether SF-MPQ-2 data (best modeled as bifactor with a strong general factor) should be
specified as a unidimensional or multidimensional measurement model in SEM (Rodriguez
et al., 2016b) [18]. These additional bifactor statistics enable researchers to determine whether
modeling multidimensional data as unidimensional in SEM would significantly bias parameter
estimates (i.e., factor loadings) (Reise et al., 2013 [46]; Rodriguez et al., 2016b [18]). The ECV
reflects the proportion of total common variance explained by the general factor, and the PUC
reflects the percentage of item correlations attributable only to the general factor. Based on
their analyses of 50 published bifactor models, Rodriguez et al. (2016a) [19] concluded that
when both the ECV and PUC are high (i.e., >.70), specifying unidimensionality in SEM is
supported. According to Rodriguez et al. (2016b) [18], ARPB values reflect “the difference
between an item’s loading in the unidimensional solution and its general factor loading in the
bifactor (i.e., truer model), divided by the general factor loading in the bifactor” (p. 145).
ARPB values less than 10–15% indicate acceptable levels of parameter bias when modeling
multidimensional data as unidimensional in SEM (Muthe
´n, Kaplan, & Hollis, 1987 [47];
Rodriguez et al., 2016b [18]). As shown in Table 3, the general pain factor accounted for over
three fourths of the total common variance (ECV = .76) and SF-MPQ-2 item correlations
(PUC = .78). Finally, ARPB values (ranging from 0–9%, M= 2%) were very low across the 22
SF-MPQ-2 items, which suggests that modeling the SF-MPQ-2 as unidimensional in SEM
would not result in significant parameter bias. Collectively, results of the reliability and
dimensionality analyses indicated strong support for the unidimensionality of the SF-MPQ-2.
Incremental validity
Finally, we used SEM to evaluate the incremental validity of the SF-MPQ-2 general and specific
pain factors in the bifactor model by examining their unique relations with five pain-related exter-
nal criteria: neck pain, depression, anxiety, stress, and symptom severity. We tested five separate
structural models, one for each of the five criteria. In each model, the SF-MPQ-2 items were mod-
eled as specified above in Model 4 (i.e., the bifactor model). We used the 10 NDI items, seven
DASS-21 depression items, seven DASS-21 anxiety items, and seven DASS-21 stress items as
observed indicators of neck pain, depression, anxiety, and stress latent variables, respectively, in
the models predicting one of those four criteria. Finally, we used raw symptom severity scores to
measure a manifest symptom severity variable in the model predicting symptom severity. In each
structural model, we regressed the criterion variable (i.e., neck pain, depression, anxiety, stress, or
symptom severity) onto the general and specific pain factors. Model fit was assessed using the
same fit indices and cutoffs as specified above in the analyses of the SF-MPQ-2’s structure.
Results revealed that all five measurement and corresponding structural models adequately
fit the data (i.e., CFI .94; RMSEA .050; SRMR <.046). As shown in Table 4, the general
Table 4. Incremental validity of SF-MPQ-2 general and specific factors (Model 4).
SF-MPQ-2 factor Neck pain Severity Depression Anxiety Stress
General Pain Factor .40*** .20*** .33*** .39*** .29***
Specific Continuous Pain Factor .13 .05 -.04 -.02 .06
Specific Intermittent Pain Factor .04 .04 .04 .03 .05
Specific Neuropathic Pain Factor .03 -.02 -.05 .00 .02
Specific Affective Pain Factor -.02 .06 .24*** .29*** .27***
Note.N= 552. SF-MPQ-2 = Short-Form McGill Pain Questionnaire-2. All values are standardized parameter
estimates (βs).
***p<.001.
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pain factor was the most consistent and strongest unique predictor of all five pain-related
external criteria. The affective pain factor uniquely and positively predicted depression (β=
.24, p<.001), anxiety (β= .29, p<.001), and stress (β= .27, p<.001). The continuous, inter-
mittent, and neuropathic pain factors did not uniquely predict any of the external criteria.
Discussion
The purpose of this study was to evaluate the structure of the widely used SF-MPQ-2 in a large
sample of adults diagnosed with Chiari malformation, a chronic health condition whose pri-
mary symptoms include pain in the head and neck region, coupled with varying degrees of
fatigue and weakness [38], depression and anxiety [29], and cognitive deficits [30–33]. We
compared CFA results of four different structural models: (a) an oblique four-factor model
based on the SF-MPQ-2’s recommended scoring (Dworkin et al., 2009) [11]; (b) a one-factor
model consistent with scoring the SF-MPQ-2 for total pain; (c) a higher-order model in which
the first-order factors of the oblique four-factor model mediate relations between a higher-
order general pain factor and SF-MPQ-2 item responses; and (d) a bifactor model in which
SF-MPQ-2 item variability is explained by both a general pain factor and specific pain factors
corresponding to the four SF-MPQ-2 subscales. Results revealed the strongest support for the
bifactor model, suggesting that SF-MPQ-2 item responses are due to both a general pain factor
and a specific pain factor that is orthogonal to the general pain factor. Additional bifactor anal-
yses of the SF-MPQ-2’s model-based reliability and dimensionality revealed that most of the
SF-MPQ-2’s reliable variance is explained by a general pain factor, and that the instrument
can be modeled unidimensionally and scored as a general pain measure. Results also indicated
that the general and affective pain factors in the bifactor model uniquely predicted pain-related
external criteria (e.g., depression, anxiety, and stress); however, the continuous, intermittent,
and neuropathic factors did not. Following is a more detailed discussion of the major findings.
SF-MPQ-2 structure
CFA results indicated that, of the four models tested, only the bifactor model adequately fit the
SF-MPQ-2 data in our sample of adults with Chiari malformation. Moreover, the bifactor
model provided a significantly better fit than did the higher-order model. These results suggest
that variability in SF-MPQ-2 item responses is attributable to a combination of a general pain
construct and specific pain constructs. These findings call into question earlier findings
regarding the structure of the SF-MPQ-2. For example, several previous CFA studies reported
support for an oblique four-factor model (e.g., Gauthier et al., 2014 [14]; Lovejoy et al., 2012
[15]) or for interpreting the four SF-MPQ-2 subscales (Dworkin et al., 2009 [11]; Dworkin
et al., 2015 [13]). However, the oblique four-factor solution in two of those studies (Gauthier
et al., 2014 [14]; Lovejoy et al., 2012 [15]) did not meet conventional fit index cutoffs for ade-
quate fit (e.g., CFI close to .95, RMSEA .06; Kline, 2016 [17]). Furthermore, Dworkin and
colleagues [11,13] conducted separate CFAs on SF-MPQ-2 item sets composing a given sub-
scale; therefore, it is unclear from their analyses to what extent the items that clustered to
reflect a narrower pain construct also (or instead) were influenced by a general source of vari-
ance (i.e., total pain).
An alternative to the oblique four-factor model—the higher-order model—also has
received support in previous research. Lovejoy et al. (2012) [15] found that a higher-order
model fit the SF-MPQ-2 data as well as an oblique four-factor model in their sample of 186 vet-
erans with chronic pain symptoms. However, neither model met recommended cutoffs for
adequate fit (e.g., CFI close to .95, RMSEA .06; Kline, 2016) [17]. Consistent with Lovejoy
et al.’s (2012) findings [15], the higher-order model fit the current data as well as the oblique
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four-factor model did; however, neither model provided an adequate fit using recommended
cutoffs.
Support for a bifactor model implies that, in addition to a general factor influencing item
responses, each subset of items corresponding to a given subscale loads substantively on (and
thus is well defined by) the corresponding specific factor (Brunner et al., 2011) [16]. In the cur-
rent bifactor model, however, only 12 of 22 SF-MPQ-2 items loaded significantly on a specific
pain factor, and only eight of the 12 significant loadings exceeded .30. In contrast, all 22 items
loaded significantly and substantively (ranging from .47 to .75) on the general pain factor. Fur-
ther, only one of the 22 SF-MPQ-2 items (item 21: “tingling”) loaded more strongly on a spe-
cific pain factor than on the general pain factor. Thus, although a multidimensional bifactor
model best captured the structure of the SF-MPQ-2 in our sample, variability in item
responses was primarily driven by the general pain factor.
Bifactor reliability and dimensionality
Consistent with recent recommendations for bifactor modeling (Rodriguez et al., 2016a,
2016b) [18,19], we calculated a number of additional bifactor statistics to evaluate the model-
based reliability and dimensionality of the SF-MPQ-2. Bifactor reliability analyses revealed
that the specific SF-MPQ-2 subscales evidenced unacceptably low reliabilities after accounting
for the variance attributable to the general pain factor (Rodriguez et al., 2016b) [18]. In con-
trast, total pain scores demonstrated a high level of reliability, suggesting that SF-MPQ-2 total
scores are mostly attributable to a general pain factor. These results support the common prac-
tice of scoring and interpreting the SF-MPQ-2 for total pain (e.g., Dworkin et al., 2015 [13];
Gauthier et al., 2104 [14]; Lovejoy et al., 2012 [15]). However, because SF-MPQ-2 raw subscale
scores primarily recapture the general pain factor instead of their intended specific pain con-
structs, interpreting the raw subscale scores as measuring unique pain constructs (i.e., beyond
general pain) is not recommended.
As further evidence of the instrument’s unidimensionality, additional bifactor indices
revealed that over three-fourths of the explained common variance and correlations among
the 22 SF-MPQ-2 items was attributable only to the general pain factor. Furthermore, although
the one-factor model provided a poor fit to the data, absolute relative parameter bias values
indicated that item factor loadings in the unidimensional solution and the general pain factor
of the bifactor solution were nearly identical. This finding suggests that although the SF-MPQ-
2 was found to be a multidimensional measure (i.e., best fitting a bifactor model), modeling
the SF-MPQ-2 items as unidimensional in SEM would not bias the instrument’s ability to cap-
ture the general pain construct of the truer (but more complex) bifactor model (Rodriguez
et al., 2016b [18]).
Incremental validity
Although our results do not support the practice of using raw SF-MPQ-2 subscale scores to
measure specific subdomains of pain, modeling the SF-MPQ-2 as a bifactor solution (i.e.,
orthogonal general and specific pain factors) in SEM revealed unique associations of general
and specific pain factors with five pain-related criteria. The general pain factor uniquely and
positively predicted neck pain, symptom severity, depression, anxiety, and stress. Beyond gen-
eral pain, the only other unique predictor of the external criteria was affective pain, which
uniquely predicted depression, anxiety, and stress. Because affective pain “refers to how
unpleasant or disturbing the pain feels” (Fillingim et al., 2016, p. T11 [7]), and past research
has reported an increase in anxiety and depression in individuals diagnosed with CM (Garcia
et al., 2019) [29], it is hardly surprising that participants’ experience of affective pain accounted
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for unique variance in their experience of depression, anxiety, and stress, all of which involve
some degree of emotional discomfort. These findings are somewhat consistent with previous
studies relating raw SF-MPQ-2 subscale scores to measures of depression and anxiety (Dwor-
kin et al., 2015 [13]; Gauthier et al., 2014 [14]; Lovejoy et al., 2012 [15]). In those studies, affec-
tive pain related to depression and anxiety only slightly more so than did the three sensory
pain constructs. It seems likely that the modest differential relations of specific pain constructs
with depression and anxiety found in previous studies were due to the raw SF-MPQ-2 subscale
scores, which primarily remeasure general pain instead of their intended specific pain con-
structs. In sum, results of incremental validity analyses suggest that, although scoring the
SF-MPQ-2 raw subscales is contraindicated, specifying a bifactor latent model in SEM with
general and specific pain constructs can allow researchers to determine the unique relations of
these constructs—most notably affective pain—to external criteria.
Implications for future research and clinical assessment
Our results of an optimal bifactor model with a general and a specific factor applied to Chiari
patients have implications for both research and application of the assessment of pain using
the SF-MPQ-2. Conceptually, the pattern of findings here might point to a general factor
attributable to the acute experience of pain associated with tissue damage, encroachment, or
inflammation (i.e., nociceptive), whereas the more specific pain factor (i.e., affective) may
reflect more chronic centralized pain (e.g., Allen et al., 2018 [20]; 2022 [25]). Evidence sup-
porting this dichotomy may be found in the pattern of less optimal post-surgical improvement
observed among Chiari patients when there is greater than a two-year interval between initial
diagnosis and surgery (Labuda et al., 2022 [48]). Such a time period might permit increased
central sensitization of pain to occur [20], quite possibly leading to increased activation of
emotive-affective networks. Also, the present incremental validity analyses underscore the
relationship between the specific (affective) factor and the DASS21 measures of Depression,
Anxiety, and Stress–variables which are routinely associated with chronic pain [29] (Garcia
et al., 2019). It is also worth noting that Gholampour and Taher (2018) [49] and Gholampour
and Gholampour (2020) [50] have implicated cerebrospinal fluid (CSF) pressure with degree
of headache pain in Chiari malformation patients (as measured by phase contrast MRI, or
cine-flow). In support of this, Garcia et al. (2022) [25] found that a smaller anterior cervical
CSF space (between C2 and the foramen magnum in the upper spine) was associated with
increased pain in Chiari patients as adult age increased. As such, future research should assess
whether CSF pressure in Chiari patients is differentially related to acute (e.g., Val Salva) and
chronic (i.e., centrally sensitized) pain.
Psychometrically, our findings also suggest that researchers can measure the SF-MPQ-2’s
general pain factor using either a bifactor model or a more parsimonious unidimensional mea-
surement model in SEM. As such, our findings support the common practice (among
researchers and clinicians) of calculating and interpreting raw SF-MPQ-2 total mean scores.
Conversely, given that very little reliable variance was explained by specific subscale scores
here, there is questionable support for using raw SF-MPQ-2 subscale scores in either a research
or clinical context. Obviously, such a finding is anticipated to raise questions and resistance,
particularly among those who have used this measure. Importantly, the intent is not to under-
mine the utility of the instrument; rather, to better understand how the SF-MPQ-2 captures
the overall experience of pain, and to align the measure with the physical, qualitative, and
affective dimensions of pain.
We recognize that our results must be considered in light of some important caveats. First
and foremost, our sample consisted of adults diagnosed with Chiari malformation, a fairly rare
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chronic health condition characterized by varying degrees of sensory, motor, cognitive and
psychosocial complications, with a higher likelihood of head and neck pain (Houston et al.,
2022 [51]; Fischbein et al., 2015 [38]). As such, we cautiously underscore our reported findings,
cognizant of the need to further assess the psychometric properties of the SF-MPQ-2 within
the broader experience of self-reported pain across diverse medical conditions. It is notewor-
thy, however, that in terms of levels of self-reported pain, our sample was quite comparable
(i.e., within roughly one-half standard deviation) to other chronic pain samples reported in the
literature (e.g., Dworkin et al., 2009 [11]; Gauthier et al., 2014 [14]; Lovejoy et al., 2012 [15]).
Nevertheless, replication with other samples representing one or more chronic (e.g., fibromy-
algia, advanced cancer, arthritis) or acute (e.g., acute low back pain) pain conditions is war-
ranted. In addition, the vast majority of our participants identified as female and white, thus
somewhat limiting the implications here. Inasmuch as women and men can differ in their sub-
jective experience of pain (e.g., Etherton, Lawson, & Graham, 2014 [52]), it remains an impor-
tant research consideration to evaluate gender effects on psychometric aspects of subjective
pain measures. Future researchers are encouraged to replicate the current study with further
attention to gender differences and sample diversity.
Conclusion
Collectively, our results provide strong support for calculating and interpreting raw SF-MPQ-
2 total scores for a general pain construct. Furthermore, because SF-MPQ-2 item responses
largely reflect a general pain construct, modeling the instrument as a unidimensional measure-
ment model in SEM is justified. Nevertheless, there may be instances when researchers wish to
examine the predictive utility of both the general and specific pain constructs. In such
instances, researchers should choose to represent the SF-MPQ-2 multidimensionally, i.e., as a
bifactor model. Finally, our results contraindicate the common practice of calculating and
interpreting the raw SF-MPQ-2 subscale scores for specific pain constructs.
Supporting information
S1 File.
(SAV)
Author Contributions
Conceptualization: David M. Tokar, Kevin P. Kaut, Philip A. Allen.
Formal analysis: David M. Tokar, Philip A. Allen.
Project administration: David M. Tokar.
Writing – original draft: David M. Tokar, Kevin P. Kaut, Philip A. Allen.
Writing – review & editing: David M. Tokar, Kevin P. Kaut, Philip A. Allen.
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