The Cervical Dystonia Impact Profile (CDIP-58): Can a Rasch developed patient reported outcome measure satisfy traditional psychometric criteria?
© Cano et al; licensee BioMed Central Ltd. 2008
Received: 08 December 2007
Accepted: 06 August 2008
Published: 06 August 2008
The United States Food and Drug Administration (FDA) are currently producing guidelines for the scientific adequacy of patient reported outcome measures (PROMs) in clinical trials, which will have implications for the selection of scales used in future clinical trials. In this study, we examine how the Cervical Dystonia Impact Profile (CDIP-58), a rigorous Rasch measurement developed neurologic PROM, stands up to traditional psychometric criteria for three reasons: 1) provide traditional psychometric evidence for the CDIP-58 in line with proposed FDA guidelines; 2) enable researchers and clinicians to compare it with existing dystonia PROMs; and 3) help researchers and clinicians bridge the knowledge gap between old and new methods of reliability and validity testing.
We evaluated traditional psychometric properties of data quality, scaling assumptions, targeting, reliability and validity in a group of 391 people with CD. The main outcome measures used were the CDIP-58, Medical Outcome Study Short Form-36, the 28-item General Health Questionnaire, and Hospital and Anxiety and Depression Scale.
A total of 391 people returned completed questionnaires (corrected response rate 87%). Analyses showed: 1) data quality was high (low missing data ≤ 4%, subscale scores could be computed for > 96% of the sample); 2) item groupings passed tests for scaling assumptions; 3) good targeting (except for the Sleep subscale, ceiling effect = 27%); 4) good reliability (Cronbach's alpha ≥ 0.92, test-retest intraclass correlations ≥ 0.83); and 5) validity was supported.
This study has shown that new psychometric methods can produce a PROM that stands up to traditional criteria and supports the clinical advantages of Rasch analysis.
Patient reported outcome measures (PROMs) are increasingly being used as primary or secondary outcome measures in clinical trials [1, 2]. As such, the quality of inferences made from clinical trials is dependent on the PROMs used. This increasingly acknowledged fact has led the United States Food and Drug Administration (FDA) to produce guidelines [3, 4] that specify minimum criteria for the scientific adequacy of scales in clinical trials. These are likely to be followed by the European Medicines Agency (EMEA) , and will have implications for all scales used in future clinical trials.
The Cervical Dystonia Impact Profile (CDIP-58) assesses the health impact of CD . It was developed using new, sophisticated, but not widely known techniques of PROMs construction (Rasch analysis), which are, as of yet, not included in the FDA guidelines. In addition, researchers interested in using the CDIP-58, who may be unfamiliar with new psychometric methods, may find the original paper  abstruse and intangible.
The aim of this study is to provide clinicians with a comprehensive evaluation of the CDIP-58 using a traditional approach to scale evaluation for three reasons: 1) provide traditional psychometric evidence for the CDIP-58 in line with the proposed FDA guidelines; 2) enable researchers and clinicians to make their own judgment of its performance and compare it with existing dystonia scales; and 3) help researchers and clinicians bridge the knowledge gap between old and new reliability and validity testing methods.
Setting and Participants
A random sample of 460 people with CD was recruited from a complete list of 1110 members from the Dystonia Society of Great Britain. The sampling strategy is described elsewhere . A booklet of questionnaires was administered as a postal survey following standard techniques . In addition, 140 individuals were randomly selected to receive a second identical battery after 2 weeks to estimate test-retest reproducibility (TRT). This study was reviewed and passed by the local hospital trust research ethics committee.
Adapted from table 4 of the FDA draft guidelines for measurement properties reviewed for PRO instruments used in clinical trials
Methods used in testing the CDIP-58
Whether the items in a domain are intercorrelated, as evidenced by an internal consistency statistic (e.g., coefficient alpha)
Inter-interviewer reproducibility (for interviewer-administered PROs only)
Agreement between responses when the PRO is administered by two or more different interviewers
Ability to measure the concept (also known as construct-related validity; can include tests for discriminant, convergent, and known-groups validity)
Whether relationships among items, domains, and concepts conform to what is predicted by the conceptual framework for the PRO instrument itself and its validation hypotheses.
Ability to predict future outcomes (also known as predictive validity)
Whether future events or status can be predicted by changes in the PRO scores
Ability to detect change
Includes calculations of effect size and standard error of measurement among others
Smallest difference that is considered clinically important; this can be a specified difference (the minimum important difference (MID)) or, in some cases, any detectable difference. The MID is used as a benchmark to interpret mean score differences between treatment arms in a clinical trial
Responder definition – used to identify responders in clinical trials for analyzing differences in the proportion of responders between treatment arms
Change in score that would be clear evidence that an individual patient experienced a treatment benefit. Can be based on experience with the measure using a distribution-based approach, a clinical or non-clinical anchor, an empirical rule, or a combination of approaches.
Data quality concerns the completeness of item- and scale-level data, and was determined by the percentage of missing data for items, and the percentage of computable scale scores . The criterion for acceptable item-level missing data was < 10%  and for computable scale scores > 50% completed items .
Items in each scale should measure a common underlying construct otherwise it is not appropriate to combine them to generate a scale score . This was evaluated by examining the correlation between each item and scale score computed from the remaining items in that scale (corrected item-total correlation). The criterion used was corrected item-total correlation ≥ 0.30 .
Items in each scale should contain a similar proportion of information concerning the construct being measured otherwise they should be given different weights . This criterion was determined by examining the equivalence of corrected item-total correlations. The criterion used was corrected item-total correlation ≥ 0.30 .
Items should be correctly grouped into scales. That is, items should correlate higher with the total score of their own scales (item own-scale correlation) than with the total score of the other scales (item other-scale correlations). The recommended criterion for definite scaling successes are item-own scale correlations (corrected for overlap) exceeding item-other scale correlations by at least two standard errors (2 × 1√n) . In situations where this criterion was not reached, we examined the magnitude of differences between item-own and item-other scale correlations. The greater the magnitude of differences between item-own scale and item other-scale correlations, the greater the support for scaling success.
The targeting of a scale to a sample indicates whether a scale is acceptable as a measure for the sample. It is recommended that: scale scores should span the entire scale range; floor (proportion of the sample at the maximum scale score) and ceiling (proportion of the sample at the minimum scale score) effects should be low (<15%) ; and skewness statistics ranging should range from -1 to +1 .
Reliability is the extent to which scale scores are dependable and consistent. Two types were examined. Internal consistency, reported as Cronbach's alpha coefficients, estimates the random error associated with scores from the intercorrelations among the items . TRT reproducibility, reported as intraclass correlations coefficients (ICC) on scores produced by a sub sample assessed twice over a 2-week interval, estimates the ability of CDIP-58 subscales to produce stable scores over a given period of time in which the respondents' condition is assumed to have remained the same . We used a two-way random effects model based on absolute agreement as a suitable, conservative estimate of test retest reliability, as this type of ICC accounts for the systematic differences among levels of ratings. This is because the raters used were only a sample of all possible raters. We used a two-way random effects model based on absolute agreement as a suitable conservative estimate test retest reliability, as this type of ICC accounts for the systematic differences among levels of ratings . Recommended criteria for adequate reliability are Cronbach's alpha coefficient ≥ 0.80 , and TRT ICC ≥ 0.80 .
Intercorrelations between CDIP-58 subscales were assessed to examine the extent to which scales measured separate but related constructs . The magnitude of intercorrelations between CDIP-58 subscale scores were predicted to be consistent with expectation about the proximity of the constructs, and were generally expected to be moderate in size (r = 0.30–0.70) . In addition, subscale reliabilities should be larger that inter-scale correlations to support that scales measure distinct constructs.
Correlations between CDIP-58 subscales and other scales were examined. Patients were asked to complete three other questionnaires for validity testing: Medical Outcome Study 36-item Short Form Health Survey (SF-36) measures health status in eight multi-item scales (Role Limitations-Emotional, Role Limitations-Physical, Bodily Pain, Vitality, General Health Perceptions, Social Functioning, Physical Functioning, Mental Health) ; 28-item version of the General Health Questionnaire (GHQ-28) measures psychological well being in four dimensions (Somatic Symptoms, Anxiety, Social, Depression) , and Hospital and Anxiety and Depression Scale (HADS) measures mood in two scales (Depression and Anxiety) . A number of hypotheses were made based on the direction, magnitude and pattern of correlations being consistent with expectations based on the proximity of the constructs.
Ideally for the results of correlations between CDIP-58 subscales and other scales to be fully interpretable the external measures should be reliable and valid. Whereas we have previously examined the psychometric properties of the SF-36 in CD , there are no current published articles which have examined the HADS or GHQ-28. Our reasoning for selecting the latter two scales was on the basis of their wide-spread use in neurologic research. Importantly, this is a common limitation of reliability and validity testing and the findings should be interpreted with this borne in mind.
The Pain and Discomfort subscale would correlate more highly with the SF-36 bodily pain than with unrelated measures of psychological functioning (SF-36 Mental Health, HADS Anxiety and Depression).
The Upper Limb and Walking subscales would correlate more highly with the SF-36 physical functioning than with unrelated measures of psychological functioning (SF-36 Mental Health, HADS Anxiety and Depression, GHQ-28).
The Annoyance and Mood subscales would correlate more highly with the SF-36 Mental Health than with unrelated measures of physical functioning (e.g. SF-36 physical functioning).
The Annoyance and Mood subscales would correlate moderately with the HADS, GHQ-28 anxiety and depression scales as these reflect aspects of mood.
The Psychosocial Functioning subscale would correlate moderately with the SF-36 social functioning as this reflects an aspect of psychosocial functioning.
Correlations between CDIP-58 subscales and sociodemographic variables (age, sex, and level of education attained) were examined to determine the extent to which they were biased by these variables. We predicted that these correlations would be low < 0.30.
Years since CD onset
2 – 50
Unable to work due to CD
External measures (Mean; SD)
SF-36** Bodily Pain
SF-36 Social Functioning
SF-36 Physical Functioning
SF-36 Mental Health
Data quality (Table 3)
Data quality, scaling assumptions, targeting, reliability and validity
Item missing data (range %)
Computable scale scores (%)
Corrected item-total correlations
Item-other scale correlations
Scaling successes (%)
Floor/ceiling effect (%)
Reliability (n = 377–385)
TRT (ICC; n = 92–95)
Data quality was high. The proportion of item-level missing data was low (≤ 4%). Subscale scores could be computed for at least 96% of the sample.
Scaling assumptions (Table 3)
Corrected item-total correlations for each of the eight CDIP subscales ranged from 0.64–0.93 satisfying the recommended criteria (> 0.30). This supported that items in each subscale of the CDIP-58 measured a common underlying construct.
Corrected item-total correlations > 0.30 indicated that items in each of the subscales contained a similar proportion of information.
Fifty-five of the fifty-eight items correlated higher with their own subscale than other subscales. Forty-seven of these exceeded the criterion (2 × 1√n). This provided some support for the grouping of items in each of the eight subscales. There was less support for three items which correlated higher with other subscales: Head Neck symptoms 'stiffness in your neck' (Pain and Discomfort subscale, r = 0.72), Pain and Discomfort 'tightness in your neck' (Head and Neck symptoms subscale, r = 0.75), and Upper Limb 'getting tired when doing demanding physical activities' (Walking subscale, r = 0.82).
Targeting (Table 3)
Subscale scores spanned the entire scale range. However, the Walking scale fell just outside of the criterion (ceiling effect = 17%) and the Sleep subscale was found to have a more significant ceiling effect (27%). Despite this, responses were not notably skewed (-0.23 to +0.82). These findings indicate good scale-to-sample targeting, thus supporting total and subscale scores as appropriate for all patients representing the full spectrum of CD impact.
Reliability (Table 3)
Cronbach's alpha, and test-retest ICCs for all eight CDIP-58 subscales were high (> 0.83), supporting their reliability.
Validity (Table 4)
Convergent and discriminant construct validity of the CDIP-58
Head and Neck Symptoms
Pain and Discomfort
Upper Limb Activities
Intercorrelations between CDIP-58 subscales ranged from 0.44 – 0.84, suggesting the subscales measured related but different constructs. A few of the correlations fell outside of the predicted range of correlations, and were highly correlated (highlighted in Table 2). However, the correlations were not unreasonable given the proximity of the constructs in each of the subscales (see Figure 1) and scale reliabilities were larger than inter-scale correlations supporting that CDIP-58 subscales measure distinct constructs.
Correlations between CDIP-58 subscales and hypothesised related scales of the SF-36, GHQ and HADS were consistent with predictions (highlighted in Table 4). This provides support that the CDIP-58 subscales measure what they intend to measure.
Correlations between CDIP-58 subscales and sociodemographic variables (age, sex, and level of education attained) were in general low (-0.17 to +0.06). This finding suggests that responses to the CDIP-58 were not biased by socio-demographic factors.
The forthcoming FDA guidelines make it increasingly important for researchers and clinicians to be exposed to the science behind PROMs. In this study, the CDIP-58 satisfied traditional psychometric criteria for data quality, scaling assumptions, targeting, reliability and validity. We hope that this, together with our previous work on conceptual model and scale development  and assessment of the sensitivity to clinical change of the CDIP-58 following Botulinum Toxin Type A (that found it to be superior to existing CD PROMs ), provides an evidence-base for its use in clinical trials, in line with the forthcoming FDA guidelines. As such, the CDIP-58 offers an advance on current PROMs. In addition, our findings are relevant to practicing neurologists, who can use this information to compare the CDIP-58 to existing published CD PROM data, which will help to avoid an ad hoc approach which may negatively impact upon rigorous measurement.
Three main issues arise from the findings. First, were there any instances where traditional psychometric criteria were not met and how should we interpret these? Second, how can the information provided here be used and what do traditional psychomteric analyses tell us? Third, what is the added value of using Rasch analysis to develop PROMs and in particular, what benefits are gained from the required additional investment in skill level, retraining and software costs?
Traditional psychometric analyses detected one problem not identified by Rasch analysis. Tests of scaling assumptions were failed by 3 items ('stiffness in the neck', 'tightness in the neck', 'upset'). This means that they correlated similarly with their own subscales and other subscales they were not intended to belong in. There are three reasons why this may be the case. First, the subscales in question were themselves highly correlated. Second, these items may be non-specific indicators of their intended construct. Third, any item can exist conceptually in more than one scale. The clinical implications of this are probably minimal, as the constructs measured by the three subscales in question are anchored by the other items, which in turn performed well psychometrically.
So, how can the information provided here be used and what do traditional psychomteric analyses tell us? Researchers who are unfamiliar with Rasch analysis can use the information presented here to compare the CDIP-58 to existing published CD PROM data. The caveat is that any inferences made from this paper alone are constrained by the sample and scale limitations inherent to all studies that use traditional psychometric analyses. These include three main points. First, total scores are often analysed as if they were interval measures. However, it has been widely demonstrated that they are not, and therefore, they are not measuring consistently across the range of the scale. Importantly, we do not know the extent to which they are measuring inconsistently across the scale. Second, traditional psychometric analyses rely directly on the items and samples used to estimate them. This means that item properties vary depending on the sample and patient scores in turn depend on the set of items taken. Thus, the reliability and validity estimates of a measure may differ across different patient groups. Third, it is recommended that total scores are only used for group comparison studies and not individual patient measurement, because the confidence intervals around individual patient scores are so wide .
Our study suggests that Rasch analysis can produce a reliable and valid measure as defined by traditional criteria. What then is the added value of using new psychometric methods? First, when scales are successfully developed using Rasch analysis it is possible to transform ordinal level scale scores into interval level measurements [29–31]. This improves the accuracy with which we can measure differences between people and clinical change. Second, Rasch analysis enables estimates suitable for individual person measurement. This can help directly inform upon patient monitoring, management and treatment for patients. Third, reliability and validity estimates computed using Rasch analysis are much less sample dependent than those derived from traditional methods. In addition, Rasch measurement methods afford more sophisticated analyses to test theoretically driven concepts and therefore provide empirical evidence for properties such as construct validity. This has important relevance for the generalisability of PROM evaluations. These benefits are further explored in other relevant articles and texts [1, 2, 32–34].
We envisage that this article, in conjunction with our previous articles on the Rasch development of the CDIP  and our recent review in Lancet Neurology  describing traditional and new psychometric techniques, can be used by researchers and clinicians to help bridge the knowledge gap between traditional and modern reliability and validity testing methods. This study has shown that Rasch analysis methods can produce a PROM that stands up to traditional psychometric criteria. A demonstration of this nature is rare. It is much more common that scales developed using traditional methods to be tested post hoc using new approaches . Nevertheless, direct comparisons of new and traditional psychometric methods of any nature in the medical literature are sparse, and at best superficial [36, 37] In part, this may be due to the fact that these two approaches cannot be compared easily as they use different methods, produce different information, and apply different criteria for success and failure. Both approaches have their supporters and traditional psychometric methods remain the dominant paradigm. However, we believe that state-of-the-art clinical trials and research would benefit from the advantages offered by Rasch analysis.
This study has shown that new psychometric methods can produce a PROM that stands up to traditional criteria and supports the clinical advantages of Rasch analysis. In addition, the CDIP-58 satisfied traditional reliability and validity criteria further supporting it as a clinically useful measure for use in routine practice, audit and treatment trials.
We wish to thank the people with CD who participated in this study and Mr Alan Leng and Ms Laura Camfield at the Dystonia Society of Great Britain for help with recruitment. This study was supported by a project grant from the Wellcome Trust. During the writing of this paper Dr Hobart benefited by being on secondment to the School of Education, Murdoch University, Perth, Western Australia, and was support by the Royal Society of Medicine (Ellison-Cliffe Travelling Fellowship), the MS Society of Great Britain and Northern Ireland, and the NHS Technology Assessment Programme (but the opinions expressed do not necessarily reflect those of the executive).
- Hobart JC: Rating scales for neurologists. Journal of Neurology, Neurosurgery, and Psychiatry 2003, 74: iv22-iv26. 10.1136/jnnp.74.suppl_4.iv22PubMed CentralView ArticlePubMedGoogle Scholar
- Hobart JC, Cano SJ, Zajicek JP, Thompson AJ: Rating scales as outcome measures for clinical trials in neurology: problems, solutions, and recommendations. Lancet Neurology 2007, 6: 1094–1105. 10.1016/S1474-4422(07)70290-9View ArticlePubMedGoogle Scholar
- Administration FD: Patient reported outcome measures: use in medical product development to support labelling claims.[http://www.fda.gov/cber/gdlns/prolbl.pdf]
- Revicki DA: FDA draft guidance and health-outcomes research. Lancet 2007, 369: 540–542. 10.1016/S0140-6736(07)60250-5View ArticlePubMedGoogle Scholar
- Agency EM: Reflection paper on the regulatory guidance for the use of the health-related quality of life (HRQL) measures in the evaluation of medicinal products. London, ; 2006.Google Scholar
- Cano SJ, Warner TT, Linacre JM, Bhatia KP, Thompson AJ, Fitzpatrick R, Hobart JC: Capturing the true burden of dystonia on patients: the Cervical Dystonia Impact Profile (CDIP-58). Neurology 2004, 63: 1629–1633.View ArticlePubMedGoogle Scholar
- Dillman DA: Mail and telephone surveys: the total design method. New York, Wiley; 1978.Google Scholar
- McHorney CA, Ware JEJ, Lu JFR, Sherbourne CD: The MOS 36-Item Short-Form Health Survey (SF-36): III. Tests of data quality, scaling assumptions and reliability across diverse patient groups. Medical Care 1994, 32: 40–66. 10.1097/00005650-199401000-00004View ArticlePubMedGoogle Scholar
- Cano SJ, Thompson AJ, Fitzpatrick R, Bhatia K, Thompson AJ, Warner TT, Hobart JC: Evidence-based guidelines for using the Short Form 36 in cervical dystonia. Movement Disorders 2006, 22: 122–127. 10.1002/mds.21187View ArticleGoogle Scholar
- Likert RA: A technique for the measurement of attitudes. Archives of Psychology 1932, 140: 5–55.Google Scholar
- Stewart AL, Ware JEJ: Measuring functioning and well-being: the Medical Outcomes Study approach. Durham, North Carolina, Duke University Press; 1992.Google Scholar
- Group TWHOQOL: The World Health Organisation Quality of Life Assessment (WHOQOL): development and general psychometric properties. Social Science and Medicine 1998, 46: 1569–1585. 10.1016/S0277-9536(98)00009-4View ArticleGoogle Scholar
- Ware JEJ, Snow KK, Kosinski M, Gandek B: SF-36 Health Survey manual and interpretation guide. Boston, Massachusetts, Nimrod Press; 1993.Google Scholar
- Hays RD, Hayashi T: Beyond internal consistency reliability: rationale and user's guide for Multi-Trait Analysis Program on the microcomputer. Behavior Research Methods, Instruments, & Computers 1990, 22: 167–175.View ArticleGoogle Scholar
- DeVellis RF: Scale development: theory and applications. In Applied social research methods. Volume 26. London, Sage publications; 1991:121.Google Scholar
- Guttman LA: Some necessary conditions for common-factor analysis. Psychometrika 1954, 19: 149–161. 10.1007/BF02289162View ArticleGoogle Scholar
- Ware JEJ, Harris WJ, Gandek B, Rogers BW, Reese PR: MAP-R for windows: multitrait /multi-item analysis program - revised user's guide. Boston, MA, Health Assessment Lab.; 1997.Google Scholar
- McHorney CA, Tarlov AR: Individual-patient monitoring in clinical practice: are available health status surveys adequate? Quality of Life Research 1995, 4: 293–307. 10.1007/BF01593882View ArticlePubMedGoogle Scholar
- Hays RD, Anderson R, Revicki DA: Psychometric considerations in evaluating health-related quality of life measures. Quality of Life Research 1993, 2: 441–449. 10.1007/BF00422218View ArticlePubMedGoogle Scholar
- Cronbach LJ: Coefficient alpha and the internal structure of tests. Psychometrika 1951, 16: 297–334. 10.1007/BF02310555View ArticleGoogle Scholar
- McGraw KO, Wong SP: Forming inferences about some intraclass correlation coefficients. Psychological Methods 1996, 1: 30–46. 10.1037/1082-989X.1.1.30View ArticleGoogle Scholar
- Nunnally JC, Bernstein IH: Psychometric theory. 3rd edition. New York, McGraw-Hill; 1994.Google Scholar
- Kaplan RM, Bush JW, Barry CC: Health status: types of validity and the index of well-being. Health Services Research 1976, 11: 478–507.PubMed CentralPubMedGoogle Scholar
- Bohrnstedt GW: Measurement. In Handbook of survey research. Edited by: Rossi PH, Wright JD and Anderson AB. New York, Academic Press; 1983:69–121.View ArticleGoogle Scholar
- Ware JEJ, Sherbourne DC: The MOS 36-Item Short-Form Health Survey (SF-36): I. Conceptual framework and item selection. Medical Care 1992, 30: 473–483. 10.1097/00005650-199206000-00002View ArticlePubMedGoogle Scholar
- Goldberg DP, Hillier VF: A scaled version of the General Health Questionnaire. Psychological Medicine 1979, 9: 139–145.View ArticlePubMedGoogle Scholar
- Zigmond AS, Snaith RP: The Hospital Anxiety and Depression Scale. Acta Psychiatrica Scandinavica 1983, 67: 361–370. 10.1111/j.1600-0447.1983.tb09716.xView ArticlePubMedGoogle Scholar
- Cano SJ, Hobart JC, Edwards M, Linacre JM, Fitzpatrick R, Bhatia K, Thompson AJ, Warner TT: CDIP-58 can measure the impact of botulinum toxin treatment in cervical dystonia. Neurology 2006, 67: 2230–2232. 10.1212/01.wnl.0000249310.25427.f2View ArticlePubMedGoogle Scholar
- Rasch G: Probabilistic models for some intelligence and attainment tests. Volume (Reprinted 1980 by University of Chicago Press, Chicago). Copenhagen Chicago, Danish Institute for Education Research; 1960.Google Scholar
- Wright BD, Masters G: Rating scale analysis: Rasch measurement. Chicago, MESA; 1982.Google Scholar
- Andrich D: Rasch models for measurement. In Sage University paper series on quantitative applications in the social sciences, 07–068. Edited by: Lewis-Beck MS. Beverley Hills, CA, Sage Publications; 1988.Google Scholar
- Wright BD: Solving measurement problems with the Rasch model. Journal of Educational Measurement 1977, 14: 97–116. 10.1111/j.1745-3984.1977.tb00031.xView ArticleGoogle Scholar
- Massof RW: The measurement of vision disability. Optometry and Vision Science 2002, 79: 516–552. 10.1097/00006324-200208000-00015View ArticlePubMedGoogle Scholar
- Andrich D: Controversy and the Rasch model: a characteristic of incompatible paradigms ? Medical Care 2004, 42: I7 - I16. 10.1097/01.mlr.0000103528.48582.7cView ArticlePubMedGoogle Scholar
- Norquist JM, Fitzpatrick R, Dawson J, Jenkinson C: Comparing alternative Rasch-based methods vs raw scores in measuring change in health. Med Care 2004, 42: I25–36. 10.1097/01.mlr.0000103530.13056.88View ArticlePubMedGoogle Scholar
- McHorney CA, Haley SM, Ware JEJ: Evaluation of the MOS SF-36 Physical Functioning Scale (PF-10): II. comparison of relative precision using Likert and Rasch scoring methods. Journal of Clinical Epidemiology 1997, 50: 451–461. 10.1016/S0895-4356(96)00424-6View ArticlePubMedGoogle Scholar
- Prieto L, Alonso J, Lamarca R: Classical test theory versus Rasch analysis for quality of life questionnaire reduction. Health Qual Life Outcomes 2003, 1: 27. 10.1186/1477-7525-1-27PubMed CentralView ArticlePubMedGoogle Scholar
This article is published under license to BioMed Central Ltd. This is an Open Access article distributed under the terms of the Creative Commons Attribution License (http://creativecommons.org/licenses/by/2.0), which permits unrestricted use, distribution, and reproduction in any medium, provided the original work is properly cited.