Methodological and Interpretive Considerations Regarding the Reported Mediation Between Patient Self-Care and Caregiver Psychological Distress [Letter]

Dear editor

We read with interest the article by Chen et al, entitled “Caregiver Burden Mediates the Association Between Patient Self-Care Behaviors and Psychological Distress in Diabetic Foot Ulcer Care,” recently published in the Journal of Multidisciplinary Healthcare.¹ The study addresses an important issue in home-based diabetic foot ulcer care by examining the relationships among patient self-care behaviors, caregiver burden, and psychological distress. The proposed role of caregiver burden as a potential pathway linking patient self-care to psychological outcomes is clinically relevant. However, several methodological and interpretive issues warrant clarification.

First, the reported caregiver composition appears difficult to reconcile with the study’s stated focus on primary family caregivers. The Methods section describes the participants as primary family caregivers and defines the primary caregiver as a family member or informal caregiver who provides the greatest amount of unpaid care. However, Table 1 categorizes the 567 caregivers as relatives (28.7%), friends (20.6%), employees/staff (27.7%), and other (22.9%). This raises an important question regarding how the eligibility criterion of “family caregiver” was operationalized. In particular, the category “employees/staff” requires clarification. Were these individuals unpaid informal caregivers, paid caregivers, or another type of care provider? This distinction is important because the caregiving relationship, financial arrangement, emotional involvement, and caregiving responsibilities may differ substantially across these groups. It also has implications for the generalizability of findings described specifically in terms of family caregiving. We therefore suggest that the authors clarify the definitions of these caregiver categories and confirm that all included participants met the stated eligibility criteria.

Second, the reported sample-size calculation does not directly justify the adequacy of the sample for the mediation analysis, which appears to be a central inferential component of the study. The authors calculated the required sample size using a formula for estimating a continuous mean, based on the standard deviation of the Zarit Burden Interview score and an allowable error, yielding a minimum sample of 499 and an adjusted target of 554 after accounting for a 10% allowance for invalid questionnaires. The final sample of 567 was subsequently considered adequate for correlation, multiple regression, and bootstrap mediation analyses. However, a precision-based calculation for estimating a mean does not establish statistical power to detect an indirect effect. Sample-size requirements for mediation depend on the magnitude of the component paths and the indirect effect, and mediation-specific or simulation-based approaches have been recommended for this purpose.^2,3 Importantly, we do not suggest that 567 participants were necessarily insufficient. Rather, the reported calculation does not directly demonstrate that the study was adequately powered for its principal mediation analysis. Providing a mediation-specific power or precision justification would therefore strengthen the statistical rationale of the study.

Third, the variance inflation factors reported in Table 3 appear difficult to reconcile with the correlation matrix reported in Table 2. The Methods section states that these diagnostics were calculated among the four core scale variables, and the correlation matrix in Table 2 is therefore sufficient to derive them. Two of the reported values, 9.625 for self-care behaviors and 1.120 for anxiety, correspond exactly to the values implied by Table 2. However, if the VIFs in Table 3 were calculated using the four-variable correlation matrix reported in Table 2, the corresponding VIFs for caregiver burden and depressive symptoms would be approximately 38.7 and 24.9, respectively, rather than the reported 8.665 and 7.943. We would therefore appreciate clarification of how the VIFs in Table 3 were calculated and whether Tables 2 and 3 are mutually consistent.

Fourth, the interpretation of the depression mediation model warrants particular attention. The correlation between caregiver burden and depressive symptoms was r = 0.979, and the adjusted model explained 96.3% of the variance in depressive symptoms. The authors appropriately acknowledge the possibility of measurement overlap, shared response tendencies, and common-method effects, and describe the depression mediation finding as hypothesis-generating rather than confirmatory; we agree with this cautious position. Mediation analysis, however, requires not only a statistically estimable indirect pathway but also a theoretically and empirically defensible distinction between the mediator and the outcome.⁴ The depression model also presents an opposite-sign pattern between the direct and indirect effects: the indirect effect through caregiver burden was −0.893, whereas the direct effect was +0.117 and remained statistically significant, so that the two components act in opposing directions relative to the total effect of −0.776. This pattern is compatible with inconsistent mediation or suppression-type effects and should not be interpreted simply as caregiver burden accounting for part of a uniformly negative association. The same direction of the direct effect appears in the anxiety model, although there it was not statistically significant (c’ = 0.019). The Conclusions nevertheless describe caregiver burden as a pathway through which the effect of patient self-care on caregiver mental health may partly operate.

The cross-sectional design further limits interpretation of the proposed sequence itself. The authors acknowledge that causal inference and temporal ordering cannot be established. This limitation affects not only causal language but also the interpretation of the indirect effect, because mediation models conceptually imply that the exposure precedes the mediator and that the mediator precedes the outcome; when all variables are measured at the same time point, this sequence cannot be empirically established.⁵ Alternative explanations therefore remain plausible. Although the proposed model suggests that better patient self-care may reduce caregiver burden and, in turn, psychological distress, greater caregiver psychological distress could also shape caregivers’ perceptions of patient self-care and their reported burden.

Overall, Chen et al provide potentially valuable evidence regarding the interrelationship between patient self-care and caregiver psychological outcomes in diabetic foot ulcer care. We believe that clarification of the caregiver composition, a mediation-specific justification of the sample size, the reported collinearity diagnostics, and the interpretation of the depression mediation model would strengthen the interpretation of the reported findings. These considerations do not diminish the clinical relevance of the study; rather, they may help distinguish the observed statistical associations from evidence supporting a specific mediation mechanism.

We appreciate the authors’ contribution to this clinically important area and hope these comments encourage further clarification and discussion regarding the interpretation of caregiver burden as a potential pathway between patient self-care and psychological distress.

Disclosure

The author reports no conflicts of interest in this communication.