The minimum clinically important difference (MCID) is used to interpret the clinical relevance of results reported by trials and meta-analyses as well as to plan sample sizes in new studies. However, there is a lack of consensus about the size of MCID in acute pain, which is a core symptom affecting patients across many clinical conditions. We identified and systematically reviewed empirical studies of MCID in acute pain. We searched PubMed, EMBASE and Cochrane Library, and included prospective studies determining MCID using a patient-reported anchor and a one-dimensional pain scale (e.g. 100 mm visual analogue scale). We summarised results and explored reasons for heterogeneity applying meta-regression, subgroup analyses and individual patient data meta-analyses. We included 37 studies (8479 patients). Thirty-five studies used a mean change approach, i.e. MCID was assessed as the mean difference in pain score among patients who reported a minimum degree of improvement, while seven studies used a threshold approach, i.e. MCID was assessed as the threshold in pain reduction associated with the best accuracy (sensitivity and specificity) for identifying improved patients. Meta-analyses found considerable heterogeneity between studies (absolute MCID: I2 = 93%, relative MCID: I2 = 75%) and results were therefore presented qualitatively, while analyses focused on exploring reasons for heterogeneity. The reported absolute MCID values ranged widely from 8 to 40 mm (standardised to a 100 mm scale) and the relative MCID values from 13% to 85%. From analyses of individual patient data (seven studies, 918 patients), we found baseline pain strongly associated with absolute, but not relative, MCID as patients with higher baseline pain needed larger pain reduction to perceive relief. Subgroup analyses showed that the definition of improved patients (one or several categories improvement or meaningful change) and the design of studies (single or multiple measurements) also influenced MCID values. The MCID in acute pain varied greatly between studies and was influenced by baseline pain, definitions of improved patients and study design. MCID is context-specific and potentially misguiding if determined, applied or interpreted inappropriately. Explicit and conscientious reflections on the choice of a reference value are required when using MCID to classify research results as clinically important or trivial.

The minimum clinically important difference (MCID) is used to interpret the relevance of treatment effects, e.g., when developing clinical guidelines, evaluating trial results or planning sample sizes. There is currently no agreement on an appropriate MCID in chronic pain and little is known about which contextual factors cause variation. This is a systematic review. We searched PubMed, EMBASE, and Cochrane Library. Eligible studies determined MCID for chronic pain based on a one-dimensional pain scale, a patient-reported transition scale of perceived improvement, and either a mean change analysis (mean difference in pain among minimally improved patients) or a threshold analysis (pain reduction associated with best sensitivity and specificity for identifying minimally improved patients). Main results were descriptively summarized due to considerable heterogeneity, which were quantified using meta-analyses and explored using subgroup analyses and metaregression. We included 66 studies (31.254 patients). Median absolute MCID was 23 mm on a 0-100 mm scale (interquartile range [IQR] 12-39) and median relative MCID was 34% (IQR 22-45) among studies using the mean change approach. In both cases, heterogeneity was very high: absolute MCID I2 = 99% and relative MCID I2 = 96%. High variation was also seen among studies using the threshold approach: median absolute MCID was 20 mm (IQR 15-30) and relative MCID was 32% (IQR 15-41). Absolute MCID was strongly associated with baseline pain, explaining approximately two-thirds of the variation, and to a lesser degree with the operational definition of minimum pain relief and clinical condition. A total of 15 clinical and methodological factors were assessed as possible causes for variation in MCID. MCID for chronic pain relief vary considerably. Baseline pain is strongly associated with absolute, but not relative, measures. To a much lesser degree, MCID is also influenced by the operational definition of relevant pain relief and possibly by clinical condition. Explicit and conscientious reflections on the choice of an MCID are required when classifying effect sizes as clinically important or trivial.

Objectives To study the impact of blinding on estimated treatment effects, and their variation between trials; differentiating between blinding of patients, healthcare providers, and observers; detection bias and performance bias; and types of outcome (the MetaBLIND study). Design Meta-epidemiological study. Data source Cochrane Database of Systematic Reviews (2013-14). Eligibility criteria for selecting studies Meta-analyses with both blinded and non-blinded trials on any topic. Review methods Blinding status was retrieved from trial publications and authors, and results retrieved automatically from the Cochrane Database of Systematic Reviews. Bayesian hierarchical models estimated the average ratio of odds ratios (ROR), and estimated the increases in heterogeneity between trials, for non-blinded trials (or of unclear status) versus blinded trials. Secondary analyses adjusted for adequacy of concealment of allocation, attrition, and trial size, and explored the association between outcome subjectivity (high, moderate, low) and average bias. An ROR lower than 1 indicated exaggerated effect estimates in trials without blinding. Results The study included 142 meta-analyses (1153 trials). The ROR for lack of blinding of patients was 0.91 (95% credible interval 0.61 to 1.34) in 18 meta-analyses with patient reported outcomes, and 0.98 (0.69 to 1.39) in 14 meta-analyses with outcomes reported by blinded observers. The ROR for lack of blinding of healthcare providers was 1.01 (0.84 to 1.19) in 29 meta-analyses with healthcare provider decision outcomes (eg, readmissions), and 0.97 (0.64 to 1.45) in 13 meta-analyses with outcomes reported by blinded patients or observers. The ROR for lack of blinding of observers was 1.01 (0.86 to 1.18) in 46 meta-analyses with subjective observer reported outcomes, with no clear impact of degree of subjectivity. Information was insufficient to determine whether lack of blinding was associated with increased heterogeneity between trials. The ROR for trials not reported as double blind versus those that were double blind was 1.02 (0.90 to 1.13) in 74 meta-analyses. Conclusion No evidence was found for an average difference in estimated treatment effect between trials with and without blinded patients, healthcare providers, or outcome assessors. These results could reflect that blinding is less important than often believed or meta-epidemiological study limitations, such as residual confounding or imprecision. At this stage, replication of this study is suggested and blinding should remain a methodological safeguard in trials.

Objectives To determine the effects of lipid based nutritional supplements with either whey or soy protein in patients with HIV during the first three months of antiretroviral treatment (ART) and to explore effects of timing by comparing supplementation at the start of ART and after three months delay. Design Randomised controlled trial. Setting Three public ART facilities in Jimma, Oromia region, Ethiopia. Participants Adults with HIV eligible for ART with body mass index (BMI) >16. Intervention Daily supplementation with 200 g (4600 kJ) of supplement containing whey or soy during either the first three or the subsequent three months of ART. Outcome measures Primary: lean body mass assessed with deuterium dilution, grip strength measured with dynamometers, and physical activity measured with accelerometer and heart rate monitors. Secondary: viral load and CD4 counts. Auxiliary: weight and CD3 and CD8 counts. Results Of 318 patients enrolled, 210 (66%) were women, mean age was 33 (SD 9), and mean BMI was 19.5 (SD 2.4). At three months, participants receiving the supplements containing whey or soy had increased their lean body mass by 0.85 kg (95% confidence interval 0.16 kg to 1.53 kg) and 0.97 kg (0.29 kg to 1.64 kg), respectively, more than controls. This was accompanied by an increased gain of grip strength of 0.68 kg (−0.11 kg to 1.46 kg) for the whey supplement group and 0.93 kg (0.16 kg to 1.70 kg) for the soy supplement group. There were no effects on physical activity. Total weight gain increased by 2.05 kg (1.12 kg to 2.99 kg) and 2.06 kg (1.14 kg to 2.97 kg) for the whey and soy groups, respectively. In addition, in the whey supplement group overall CD3 counts improved by 150 cells/µL (24 to 275 cells/µL), of which 112 cells/µL (15 to 209 cells/µL) were CD8 and 25 cells/µL (−2 to 53 cells/µL) were CD4. Effects of the soy containing supplement on immune recovery were not significant. The effects of the two supplements, however, were not significantly different in direct comparison. Exploratory analysis showed that relatively more lean body mass was gained by patients with undetectable viral load at three months. Patients receiving delayed supplementation had higher weight gain but lower gains in functional outcomes. Conclusions Lipid based nutritional supplements improved gain of weight, lean body mass, and grip strength in patients with HIV starting ART. Supplements containing whey were associated with improved immune recovery. Trial registration Controlled-trials.com ISRCTN32453477.

The burden of diabetes is increasing in sub-Saharan Africa, including among people living with HIV. We assessed the prevalence of diabetes and the roles of HIV, antiretroviral therapy (ART) and traditional risk factors among adults in Tanzania. We analysed diabetes-relevant baseline data from 1,947 adult participants in the CICADA study in Mwanza, Tanzania: 655 HIV-uninfected, 956 HIV-infected ART-naïve, and 336 HIV-infected persons on ART. WHO guidelines for haemoglobin A1c (HbA1c) and oral glucose tolerance test (OGTT) were used to define diabetes and prediabetes. Risk factors were evaluated using multinomial logistic regression analysis. Relative risk ratios (RRR) were generated comparing participants with diabetes and prediabetes against the reference of those with no diabetes. Mean age was 41 (SD 12) years; 59% were women. The prevalence of diabetes was 13% by HbA1c and 6% by OGTT, with partial overlap among participants identified by the two tests. Relative to HIV-uninfected, HIV-infected ART-naïve persons had increased relative risks of diabetes (HbA1c: RRR = 1.95, 95% CI 1.25–3.03; OGTT: RRR = 1.90, 95% CI 0.96–3.73) and prediabetes (HbA1c: RRR = 2.89, 95% CI 1.93–4.34; OGTT: RRR = 1.61, 95% CI 1.22–2.13). HIV-infected participants on ART showed increased risk of prediabetes (RRR 1.80, 95% CI 1.09, 2.94) by HbA1c, but not diabetes. CD4 count < 200 cell/μL at recruitment increased risk and physical activity decreased risk of diabetes by both HbA1c and OGTT. The prevalence of diabetes was high, especially among HIV-infected ART-naïve adults. Being more physically active was associated with lower risk of diabetes. HbA1c and OGTT identified different participants as having diabetes or prediabetes. Overall, the finding of high burden of diabetes among HIV-infected persons suggests that health systems should consider integrating diabetes screening and treatment in HIV clinics to optimize the care of HIV patients and improve their health outcomes.