Lateral ankle sprains are highly prevalent and result in tissue damage, impairments of muscle strength, instability, and muscle activation. Up to 74% will experience ongoing symptoms after a lateral ankle sprain. In healthy subjects, motor imagery might induce neural changes in the somatosensory and motor areas of the brain, yielding favourable enhancements in muscular force. However, during motor imagery, difficulties in building a motor image, no somatosensory feedback, and the absence of structural changes at the level of the muscle might explain the differences found between motor imagery and physical practice. In rehabilitation, motor imagery might be supportive in rebuilding motor networks or creating new networks to restore impairments in muscle activation and movement patterns. This systematic review was undertaken to summarize the current body of evidence about the effect on motor imagery, or action observation, on lower leg strength, muscle performance, ankle range of motion, balance, and edema in persons with, and without, a lateral ankle sprain compared to usual care, a placebo intervention, or no intervention.th June 2023 were included. Two reviewers individually screened titles and abstracts for relevancy using the inclusion criteria. Variables related to muscle strength, muscle function, range of motion, balance, return to sports tests, or questionnaires on self-reported function or activities were extracted. A risk of bias assessment was done using the Cochrane Risk-of-Bias tool II by two reviewers. Meta-analysis using a random effects model was performed when two or more studies reported the same outcome measures. The Standardized Mean Difference (SMD) was calculated over the change from baseline scores. Review manager 5.4 was used to perform analysis of subgroup differences and test for statistically significant differences. Confidence intervals were visually checked for overlap between subgroups.A systematic review with meta-analysis of randomized controlled trials was conducted in healthy participants and participants with a lateral ankle sprain. Motor imagery or action observation in isolation, or in combination with usual care were compared to a placebo intervention, or no intervention. An electronic search of MEDLINE, EMBASE, Cinahl, Psychinfo, Sportdiscus, Web of Science, Cochrane and Google Scholar was conducted, and articles published up to 72 = 73%), limitations in the studies , and imprecision (n =  < 300). Evidence showed no association with ankle range of motion , edema , the anterior reach direction of the Star Excursion Balance Test (SEBT) , the posterolateral direction , and the posteromedial direction . The certainty of evidence for the different comparisons was very low.Nine studies, six examining healthy participants and three examining participants with an acute lateral ankle sprain, were included. All studies were rated with moderate to high risk of bias overall. Quality of the motor imagery interventions differed largely between studies. Meta-analysis showed a large and significant effect of motor imagery on lower leg strength ; however, the evidence was downgraded to very low certainty due to substantial heterogeneity (IThere is a low certainty, significant, positive effect for motor imagery being able to improve lower leg muscle strength in healthy participants. The effect on balance, range of motion and edema was uncertain and of very low certainty.PROSPERO CRD42021243258.The online version contains supplementary material available at 10.1186/s12891-023-06906-9. Lateral ankle sprains are highly prevalent in the general population and during sports activities , 2. It iMovement representation techniques can be defined as “the representation of movement, especially observation and/or imagination of normal pain-free movement” . Motor iIn rehabilitation, motor imagery might be supportive in rebuilding motor networks or creating new networks to restore impairments in muscle activation and movement patterns  and is oOver the last twenty years, several studies have looked at the effect of motor imagery on ankle function and performance in healthy participants , 29 and What is the effect of motor imagery or action observation on ankle function  and performance  in persons with, or without, a lateral ankle sprain?This systematic review was performed in accordance with the guideline Preferred Reporting Items for Systematic reviews and Meta-Analysis: The PRISMA statement  and was th June 2023 were included in the search. The results were uploaded to Rayyan . Articles published up to 7o Rayyan , and dupMedical Subject Headings (MeSH) and free-text words were used to design the search string. Boolean operator “AND” was used to combine keywords and “OR” was used to combine the keywords within each group. The following search terms were used and combined to search the databases: motor, mental, movement, mirror, imagery, ankle, injuries. The final searches for all databases can be found in Appendix If disagreement upon article relevancy existed between the two reviewers, consensus was achieved by discussion. Full texts of eligible articles were retrieved and evaluated individually by the two reviewers. Again, disagreements were resolved using discussion to reach consensus. A third reviewer [HN] was consulted to make the final decision if disagreement remained during the process. Additionally, reference lists of included articles were checked manually to identify additional literature. The Kappa statistic and percentage of agreement as a measure for interrater reliability were calculated and Kappa results were interpreted as: 0.01–0.20 none to slight, 0.21–0.40 fair 0.41–0.60 moderate, 0.61–0.80 substantial, and 0.81–1.00 almost perfect agreement.A risk of bias assessment was conducted blinded and individually by two reviewers [LS and MW]. The Cochrane risk-of-bias tool 2 for randomized trials was used to judge the domains: bias arising from the randomization process; bias due to deviations from intended interventions; bias due to missing outcome data; bias in measurement of the outcome and bias in selection of the reported result . Upon diThe data collection process was performed using a data extraction form . The ite2 as it is preferred over the Chi-square test and was considered not important if I2 was between 0 and 30%; moderate between 30 and 50%, substantial between 50 and 75%, and considerable above 75% [Meta-analysis using a random effects model was performed when two, or more, studies reported the same outcome measures in a comparable study population. Statistical heterogeneity of the intervention effect was assessed using the Ibove 75% . Review bove 75%  and SPSSbove 75% , authorsThe trial registers ClinicalTrials.gov, EU Clinical Trials Register, Netherlands Trial Register, WHO International Clinical Trials Registry Platform (ICTRP) and ISRCTN registry were checked to identify unpublished trials.The qualitative analysis of the body of evidence was performed using the Grading of Recommendations, Assessment, Development and Evaluation (GRADE) framework. The certainty of evidence was classified as high, moderate, low, or very low certainty. Evidence from randomized controlled trials began as high certainty evidence yet could be downgraded based on concerns in any of the following five categories: risk of bias, imprecision, inconsistency, indirectness, and publication bias –45.The quality of the motor imagery interventions was analysed using fifteen predefined criteria by Schuster et al. . Motor iMotor imagery sessions were scored successful when: (1) performed individually; (2) added after physical practice; (3) were supervised; (4) not directed; (5) the location of the motor imagery and (6) position of the participants was task-specific; (7) accompanied by acoustic and (8) detailed MI instructions; (9) performed with the eyes closed; (10) the perspective used during MI was an internal view with (11) kinaesthetic mode and (12) MI interventions included primarily motor-focused activities. (13) The average duration of a study was around 34 days, (14) with a total of 3 MI training sessions per week, and (15) had an average duration of 17 min per training session (with a minimum of 178 min) .n = 3) [n = 5), [n = 2; Arabic, n = 2) [n = 1) [An electronic search of the databases identified 743 records  , 48, 49,[n = 5), –54 langu, n = 2) –58, or p [n = 1) . The Kap [n = 1) , 60–64. Risk of bias assessment revealed at least some concerns in six studies , 63, 64 n =  < 10) included per meta-analysis [One trial protocol was identified, no further (published) pre-analysis plans were discovered, increasing risk of bias in the ‘selection of the reported result’ domain . Five tranalysis .n = 400; 25 players per squad). The article does not provide detailed information about the intervention in both groups, the outcome measures and which measurement instruments were used. Data extraction was limited to participant information. Most studies divided participants into an intervention group and a control group, except for two studies [The characteristics of the nine included studies can be found in Table  studies , 64, one studies  and one  studies  participIn the studies using injured athletes, the control groups received normal physiotherapy and the intervention groups received normal physiotherapy and an additional motor imagery intervention –32. The An assessment of the quality of the motor imagery intervention based on the criteria of Schuster et al.  can be fThree studies used the Movement Imagery Questionnaire-Revised (MIQ-R) and one the Movement Imagery Questionnaire-Revised second version (MIQ-RS) to assess the individual’s ability to image four movements using internal visual imagery, and kinaesthetic imagery , 63, 64.Grosprêtre et al.  reportedIn the two studies of Christakou et al. , 32 the Meta-analysis was performed for studies that compared motor imagery to control groups for lower leg strength, ankle range of motion, edema, and balance. One author  was cont2: 73%; p = 0.005). The evidence was downgraded to very low certainty due to substantial heterogeneity (I2 = 73%), limitations in the studies , and imprecision (n =  < 300)  Table .2: 41%; p = 0.47). The evidence was downgraded to very low certainty due to moderate heterogeneity (I2 = 42%), limitations in the studies (high risk of bias in one study and some concerns for risk of bias in two studies), indirectness (due to differences in study populations and interventions), and imprecision (n =  < 300).Meta-analysis of three studies Fig.  with a t2: 91%; p = 0.42). The evidence was downgraded to very low certainty due to considerable heterogeneity (I2 = 91%), limitations in the two studies (high risk of bias in one study and some concerns for risk of bias in the other study), indirectness (due to differences in the intervention), and imprecision (n =  < 300).Meta-analysis of two studies Fig.  with a t2: 79%; p = 0.29), very low certainty evidence of a small, non-significant effect of motor imagery on the posterolateral direction   , indirectness (due to differences in the study population and intervention), and considerable heterogeneity for the anterior and posterolateral direction (I2 = 79% and 76%).Meta-analysis of two studies Fig. a with a This systematic review with meta-analysis provides an overview of the evidence on the effect of motor imagery on ankle strength, range of motion, balance, and edema in persons with, or without, a lateral ankle sprain. Very low-certainty evidence for a significant, positive effect for motor imagery being able to improve lower leg muscle strength in healthy participants was found. The evidence for balance, ankle range of motion and edema in healthy and injured participants was uncertain, and of very low certainty.Several methodological strengths for this systematic review can be defined: the protocol was designed using the Cochrane Handbook for Systematic Reviews  and Prefn = 9) with heterogeneous backgrounds resulting in a limited number of studies and participants per outcome measure. The small number of studies could have underpowered the results. Moreover, certainty of evidence was low to very low because of moderate to high risk-of-bias in the included studies. The risk-of bias assessment showed problems in the blinding of the assessors performing the pre- and post-measurements which could have led to performance bias. Further, no placebo interventions were given to the control groups. A well-developed placebo is hard to construct in motor imagery study designs, however, several published randomized controlled trials in other areas than the ankle joint have used a placebo intervention [This study has several limitations. A weakness of the current systematic review and meta-analysis is the small number of studies  revealed a large, significant effect for motor imagery being able to improve lower leg strength in healthy participants , i.e., the plantar and dorsal flexor muscles. The quality of the motor imagery intervention according to criteria of Schuster et al. [Although caution must be exercised in interpreting the results of the current meta-analysis, the pooled results  of the meta-analysis for ankle range of motion revealed no association with  motor imagery improving range of motion of the ankle compared to a control group. An important difference between the three studies can be found in the content of the motor imagery interventions. Especially, the study of Nunes et al. [The pooled results  showed no association, with only a trend for the posteromedial direction on the Star Balance Excursion Test. When comparing the motor imagery interventions between the two included studies in the meta-analysis for balance, heterogeneity is observed. Nagar & Noohu [With respect to balance, the body of evidence , no association with motor imagery was found . The results were classified as indicating very low- certainty evidence. Only two studies could be included in the meta-analysis, again the large effect in favour of the control group in the study of Nunes et al. [Regarding edema (s et al.  had a stDue to the very low certainty of the evidence, the effect of motor imagery on muscle strength, ankle range of motion, balance, and edema, in persons with and without a lateral ankle sprain is still uncertain. It is recommended that researchers undertake more high-quality studies with larger sample sizes. Studies should use a randomized controlled trial design with blinded assessors during pre- and post-intervention measurements to decrease the changes of performance bias. Placebo motor imagery intervention should be developed to at least blind the participants, and the motor imagery intervention should score positive on as many of the 15 criteria of Schuster et al. as possible . The criAdditional file 1. Supplementary material.


Postoperative radiotherapy (PORT) reduces local failure in patients with NSCLC, without a clear overall survival benefit. It is unknown whether the subsets of patients benefit. Two recent large randomized controlled trials, PORT-C (People’s Republic of China) and Lung ART (Europe), reported widely different locoregional recurrence (LR) rates in the control arms, at 18.3% and 28.1% , respectively. We performed a meta-analysis of patients with pathologic (p) N0 to N2 disease to evaluate the risk factors for LR and to explore possible differences in recurrence risk between Asian population (AP) and non-Asian population (NAP).We identified all original studies of curative NSCLC surgical resection which reported risk of LR between January 1, 2000, and January 10, 2021, excluding studies with less than 10 LR, patients with metastatic disease, or any neoadjuvant therapy. A total of 87 studies were identified with pN0 to N2 disease; of these, 56 were of high quality (HQ) on the basis of the Newcastle-Ottawa Scale. For each risk factor, we derived pooled relative risk (RR) and 5-year rate estimates using random-effects models.Overall, the three significant highest pooled RRs  for LR were pN2 versus pN0 , lymphovascular invasion , and advanced pT3–4 stage versus pT1 . For HQ studies, the highest RRs for LR were lymphovascular invasion , sublobar versus lobar resection , and pN1 versus pN0 , but pN2 versus pN0 was no longer significant , on the basis of only two eligible studies. The RRs for LR were consistent for most factors in AP and NAP, although the RR for male versus female sex was higher in AP  than in NAP . Where reported, the pooled rate of LR at 5 years was lower in AP (12.0%) than in NAP (22.7%), despite similar overall 5-year recurrence rates  in both populations: 38.0% in AP and 37.3% in NAP. Nevertheless, a lower 5-year mortality rate was noted in AP (24.3%) than in NAP (45.9%).There is little high-quality evidence to support the hypothesis that pN2 disease is a risk factor for LR, but LR seems to be lower in Asians. Prospective evaluation of LR factors and rates may be necessary before further prospective evaluation of PORT, because it may not depend on nodal status alone. Recurrence rates may differ in Asians. The impact of mutational status and modern treatment including targeted therapies and immune checkpoint inhibitors is inadequately studied. We could not evaluate the impact of driver mutations as these were not performed in most studies.,The present systematic review and meta-analysis was conducted according to the Preferred Reporting Items for Systematic Review and Meta-Analysis guidelines,In the present meta-analysis, we included original study publications that satisfied the following eligibility criteria: (1) provided data on patients with stages I to III NSCLC treated with surgical resection for curative intent; (2) were prospective or retrospective observational studies or trials; (3) considered the following major risk factors for LR: sex, age, smoking status, adjuvant chemotherapy, type of resection , video-assisted thoracoscopic surgery, number of lymph node resected, lymphovascular invasion (LVI), visceral pleural invasion (VPI), pathologic stage, N stage, T stage, tumor grade, tumor histology , tumor location, and tumor size; (4) noted information on LR patterns, including rates of LR based on the factors of interest and an estimate of the relative risk , and corresponding 95% confidence interval (CI) or sufficient information was provided to compute them; and (5) were published as original articles in English. Studies were excluded if: (1) the patient number or LR number was less than 10; (2) studies included inoperable or metastatic patients; (3) any neoadjuvant chemotherapy or any radiation was given; or (4) information on risk factors listed previously was not provided.Two reviewers independently screened the titles or abstracts of the identified records to exclude publications that did not meet the eligibility criteria. Subsequently, full-text articles were retrieved and assessed for inclusion in the systematic review. Any disagreement was resolved by consensus among the two reviewers or with the help of a third reviewer. It should be noted that one of 548 patients and two of 261 patients in the manuscripts by Kim et al.We evaluated the quality of the studies using the Newcastle-Ottawa scale (NOSc)p values.For each identified study, we collected the following information: first author, publication year, study design, country, study period, total number of patients, study population, TNM edition, resection type, mean follow-up duration, end point , recurrence definition, diagnostic methods for recurrence, and numbers of LR or DR recurrence. For each risk factor investigated, we extracted data on HRs, ORs or survival rates, and corresponding 95% CIs (or information to derive them) for LR and, for comparative purposes, DR, OvRec, and OS. Whenever available, we retrieved estimates adjusted for multiple potential confounding factors. When the HRs, ORs, or corresponding 95% CIs were not reported, an estimate of the RR was computed from raw tabular data, survival rates, and corresponding ,For each risk factor and outcome of interest, we derived pooled RR estimates for exposed versus not exposed (or for different categories of exposures) using random effects models, to take into account the heterogeneity of risk estimates.p < 0.10) and quantified the inconsistencies using the I2 statistic, which expresses the percentage of total observed variability owing to heterogeneity.,We assessed heterogeneity between studies using Cochran’s chi-square test .Overall, 1632 studies were screened on the basis of the abstract and title and 100 articles were considered for inclusion in the analysis . We exclThe pooled RRs for LR are outlined in p = 0.003). In addition, age as a continuous variable was found to be of borderline significance for LR in the AP and significantly different as compared with the NAP. N2 versus N0 was significant only in the AP and significantly different than the NAP, but these findings are based only on one and two studies in the AP and NAP, respectively.The pooled rate of LR at 5 years was lower in the AP  versus NAP . Stage had mostly remained significant for DR in both populations, but in stages Ib to II versus Ia, the RR was higher in the NAP than in the AP .The pooled rate of DR at 5 years was 22.4% (16.02–30.72) in the NAP, whereas none of the AP publications reported data on pooled DR (data not found). p = 0.024). Ever smoking was only significant for OvRec in the AP and significantly different than the NAP . LVI was significant in both populations for OvRec, but significantly more so in the AP . Moderate/poor tumor grade was significantly associated with DR in both populations, but it was significantly greater in the AP than the NAP . All cases of node involvement were significantly associated with OvRec in both populations, but more so in the AP than NAP for all categories, respectively, as follows: N1 versus N0 2.05 (1.58–2.67) versus 1.54 , N2 versus N0 2.58 (2.24–3.63) versus 1.74 , and N1 to N2 2.71 (2.37–3.10) versus 1.86 .The 5-year OvRec was similar in both populations: 38.0% (25.2–50.9) in AP and 37.3% (32.4–42.2) in NAP (data not found). The difference between both populations for OvRec factors is displayed in p = 0.014; moderate/poor differentiation: 3.98, 2.13–7.44, versus 1.46, 1.21–1.76, p = 0.003) (p = 0.033). Similarly, the effects of N1 and N2 on OS were significant in both populations, but more so in the AP than in the NAP .A lower 5-year mortality rate was noted in AP  than in NAP . The OS was significantly affected by LVI and moderate/poor differentiation, but the effect was significantly greater in the AP than the NAP  . Nonaden,EGFR mutations occur far more often in nonsmoking Asian females and the natural history of this entity is unique, we surmised that these and other differences could drive differences in local relapse and the underlying risk factors. Although it is difficult to compare different trials, it seems that LR was higher in the control arm of the Lung ART trial, which noted that 46% of recurrences were in the mediastinum.EGFR mutation (EGFRmut) frequency, which were found to be present in 47% (range 20%–76%) of adenocarcinomas in the AP as compared with 15%,We decided to perform this meta-analysis to clarify the risk factors for LR because of differences in the LR noted in both control arms of the recently published Lung ART and PORT-C trials in European and Chinese populations, respectively.7% range %–76% of 7% range %–76% of The main risk factors for LR in our meta-analysis were N2, LVI, and advanced T stage (T3–T4). When including only HQ studies by the NOSc, the highest RRs for LR were LVI, sublobar resection, and N1, but N2 was no longer significant, with the caveat that this conclusion is based only on two studies. The present meta-analysis thus reveals the limited information in the clinical literature pertaining to nodal involvement and its relationship to recurrence patterns. Only three and nine studies of 87 eligible studies had information pertaining to N2 and N1 nodal disease, respectively, which fell to two and eight studies when considering HQ studies. Although the RR for N2 disease remained high for LR using HQ studies, it should be noted that when combining N2 with N1 disease as compared with N0 nodal status, RRs for LR were lower in all combined studies  and in the HQ study population as compared with the RRs for N1 nodal status versus N0 . We speculate based on these data that perhaps with a larger number of studies, N2 nodal status may not be as important for LR as N1 nodal status. One previous study hypothesized that N1 rather than N2 nodal involvement may be more likely to result in LR because N1 disease is less likely to fail distally.Although the RR for LRs in patients with pN2 disease in the AP  is much higher than that in the NAP , this is a RR and not absolute risk for LR. This effect is largely because of the low rate of LR in the AP patients with resected N0 disease. In fact, the rate of LRs at 5 years in patients with pN0 disease was higher in NAP patients (14.9%)It is interesting to note that adjuvant chemotherapy had no effect on LR in the total population nor in the HQ study subset. Nevertheless, the influence of systemic therapy on recurrence patterns may change now that adjuvant chemo/immunotherapy in programmed death-ligand 1 (PD-L1)–positive NSCLC,,,,Considering only HQ studies, it is not surprising that sublobar resection, whether a known wedge resection/segmentectomy or a sublobar resection NOS, was associated with a statistically significant risk for LR. Both published prospective, randomized trialsThe pooled 5-year rate of LR was lower in the AP versus NAP, although this is based on only two AP studies reporting 5-year LR rates. The RRs for LR were consistent for most factors in AP and NAP, although the RR for male versus female sex was higher in AP than in NAP. Risk factors for LR noted in both populations are LVI, VPI, nonadenocarcinoma, and tumor size with all risk factors associated with a greater RR on LR in the AP. Although the staging variables had general trends toward an increased risk of LR, each factor is associated with a scanty number of articles in both populations except the node-positive category (N1–N2 versus N0) which reveals that the effect in the AP is not significant . Because factors such as LVI , VPI , and tumor size  were associated with a higher and significant RR for LR in the AP, this may indicate why the PORT-C trial, which randomized to PORT or not without stratification factors, represents an underpowered positive trial for the beneficial effects of PORT. The effect of nodal involvement in the NAP is not only significant , but, with the exception of T3 to T4 tumors, the RR is much higher than large tumor size , poor differentiation , or pleural invasion , and possibly explains why the mediastinal failure rate of 28.1% was much higher in the Lung ART trial.Although similar 5-year OvRec rates were noted in both the AP and NAP, survival was lower in the NAP. Of note, most factors that were significantly associated with both DR and OvRec in both populations such as tumor size, LVI, VPI, and poor differentiation are considered high-risk factors warranting adjuvant systemic therapy in node-negative resected NSCLC,,47One can question the validity of the present meta-analysis because it includes all eligible surgical series irrespective of the surgical quality. For example, one recent investigation assessed eight sequentially more stringent definitions of pathologic node evaluation and revealed that the 5-year survival was sequentially improved for N0 and N1 tumors. In addition, this study noted that the more stringent definitions of mediastinal node examination were associated with a larger separation of survival curves between N1 and N2 tumors.,It should be noted that this meta-analysis, although thorough and comprehensive, is limited by the methods of detection/surveillance and retrospective design of many included studies. The definition of LR differed between studies and there was no way ex post facto for controlling for the differences. This definition even differs between large surgical groups.,A major impetus for performing this research study was to inform lung cancer researchers that before initiation of the PORT-CIn conclusion, we revealed that the AP may be at a lower risk of local recurrence compared with NAP and that N2 nodal disease may not be the best and only determinant for risk of locoregional failure in surgically resected NSCLC. The lower risk of LR in the AP may have led to an underpowered trial, the phase 3 PORT-C trial, to determine the impact of radiation in the N2 population. Now that we have found that PORT can be given with no therapy-related grade 4 or 5 toxicityJohn M. Varlotto: Conceptualization, Investigation, Methodology, Writing .Cristina Bosetti: Data curation, Formal analysis, Investigation, Methodology, Writing .Dwight Bronson: Conceptualization, Funding Acquisition, Investigation, Methodology, Writing .Claudia Santucci: Data curation, Formal analysis, Investigation, Methodology, Writing .Maria Vitttoria Chiaruttini: Data curation, Formal analysis, Investigation, Methodology, Writing .Marco Scardapane: Investigation, Methodology, Writing .Minesh Mehta: Investigation, Methodology, Writing .David Harpole: Investigation, Methodology, Writing .Raymond Osarogiagbon: Investigation, Methodology, Writing .Gerald Hodgkinson: Conceptualization, Funding Acquisition, Investigation, Methodology, Writing .


MTHFR) gene polymorphisms and GC risk, we observed inconsistent results, and the credibility of the significant correlation between the statistical results was ignored. With the aim of further exploring the association between MTHFR C677T and A1298C and the risk of GC, we searched electronic databases, pooling 43 relevant studies and calculating odds ratios (ORs) and corresponding 95% confidence intervals (CIs) for each of the five genetic models. Subgroup and regression analyses were performed to look for sources of heterogeneity and publication bias was assessed by funnel plots. To assess the plausibility of statistically significant associations, we used the FPRP test and the Venice criteria. Overall data analysis showed that MTHFR C677T polymorphism was significantly associated with GC risk, especially in Asians, while MTHFR A1298C polymorphism was not associated with GC risk. However, in subgroup analysis by hospital-based controls, we found that MTHFR A1298C might be a protective factor for GC. After credibility assessment, the statistical association between MTHFR C677T and GC susceptibility study was classified as ‘less credible positive result’, while the result of MTHFR A1298C was considered unreliable. In summary, the present study strongly suggests that MTHFR C677T and A1298C polymorphisms are not significantly associated with the GC risk.Widely regarded as one of the most prevalent malignancies worldwide, gastric cancer (GC) is a common clinical condition of the digestive system. Reviewing 14 meta-analyses that evaluated the association between methylenetetrahydrofolate reductase ( MTHFR gene polymorphisms and GC ; (4) recessive model (C677T: TT vs. (CT + CC); A1298C: CC vs. (AC + AA); (5) over-dominant model .We used Chi-square based Q-test and I-square test for heterogeneity to assess whether heterogeneity exists between the included literature. If P>0.10 and/or I-square ≤ 50%, we considered no significant heterogeneity among studies [P>0.05 which were defined as HWE, otherwise Hardy–Weinberg disequilibrium (HWD) in control groups. Only high-quality and HWE-compliant studies were used in sensitivity analyses based on the quality score results and HWE conditions. Sensitivity analyses were made through the following three methods: (1) by sequentially excluding one study, (2) by excluding low- and moderate-quality or HWD studies, and (3) by retaining only high-quality and HWE studies. In the meantime, the false positive reporting probability (FPRP) test [By calculating the pooled odds ratios (ORs) with corresponding 95% confidence intervals (95% CIs) for each genetic model gene frequency,  studies  and pool studies . If not, studies  and usedRP) test  and the RP) test , we applRP) test  and EggeRP) test . All staMTHFR C677T)/random effects model (MTHFR A1298C) as in the previous meta-analysis [The trial sequential analysis (TSA) is performed as we previously described , in shoranalysis , and typanalysis .Click here for additional data file.


The PRISMA statement was strictly followed, and the protocol was registered in PROSPERO (CRD42022339317). The PICOS framework was used: patients received antituberculosis treatment, UGTs polymorphisms (mutants), UGTs polymorphisms (wild), AT-DILI, and case-control studies. Eligible studies were searched through nine databases up to April 27, 2022. The study's qualities were assessed by the revised Little's recommendations. Meta-analysis was conducted with a random-effects model using odds ratios (ORs) with 95% confidence intervals (95% CIs) as the effect size. P = 0.009), UGT2B7 rs7662029 , and UGT2B7 rs7439366  might be associated with the risk of AT-DILI.  Twelve case-control studies with 2128 cases and 4338 controls were included, and 32 single nucleotide polymorphisms (SNPs) in the seven UGT genes have been reported in Chinese and Korean. All studies were judged as high quality. The pooled results indicated that UGT1A1 rs3755319 (AC vs. AA, OR = 1.454, 95% CI: 1.100–1.921,  The polymorphisms of UGT1A1 rs3755319, UGT2B7 rs7662029, and UGT2B7 rs7439366 were significantly associated with AT-DILI susceptibility. However, this conclusion should be interpreted with caution due to the low number of studies and the relatively small sample size. Mycobacterium tuberculosis that contributes to high morbidity and mortality worldwide. In 2021, an estimated 10.6 million people fell ill with TB worldwide [Tuberculosis (TB) is a chronic communicable disease caused by orldwide . The reporldwide . A signiorldwide . The cliorldwide , which corldwide . TherefoUntil now, the specific mechanisms associated with AT-DILI have been inadequately described. A number of hypotheses on the pathogenesis of AT-DILI have been proposed, such as drug metabolism and transport, immune response, oxidative stress, and mitochondrial dysfunction . Many prHuman UGT genes have a large number of genetic polymorphisms, which have been confirmed to modulate enzymatic activity or promoter activity , furtherThis systematic review and meta-analysis was reported according to the PRISMA guidelines  and has Records identified through all searches were imported into the EndNote X9 software for screening studies, and duplicate records were removed. Two reviewers independently selected the relevant studies meeting the eligibility criteria by titles and abstracts. Full texts were referred to when the above information was inadequate or unspecific for the determination of eligibility. Any disagreement was discussed and resolved by consensus or by consulting a third reviewer.The studies included in this research met the following eligibility criteria: (1) case-control studies designed to investigate the relationship between UGTs polymorphisms and AT-DILI; (2) all patients received anti-TB treatment, of which the case group had AT-DILI, while the control group did not have AT-DILI; (3) the genotype frequency data could be extracted and analyzed; and (4) the language was restricted to English or Chinese.The exclusion criteria were the following: (1) conference abstracts, editorials, letters, case reports, reviews, and meta-analyses; (2) sample size for each group of less than 10; and (3) studies with repetitious data (the studies with the most recent or comprehensive data were selected).The following data were extracted from all included studies: (1) basic characteristics: the first author, publication year, and country of origin; (2) study characteristics: study design, sample size, diagnostic criteria of AT-DILI, method of causality assessment, treatment regimens, and genotyping method; (3) population characteristics: sex and mean age of total subjects; and (4) polymorphism results: genotype frequencies in AT-DILI cases and controls or adjusted odds ratios (ORs) with 95% confidence intervals (CIs) under different genetic models and covariates. The data extraction procedure was also performed independently by two reviewers. If there was any disagreement, it was resolved by a third reviewer. No authors were contacted for further information.The study qualities were assessed by the revised Little's recommendations . These cZ test and Mantel–Haenszel random effects model, with P < 0.05 being considered statistically significant. The heterogeneity between studies was quantified by the Cochran Q test and the I2 statistic  [The ORs and corresponding 95% CIs were calculated to identify the potential association between susceptibility to AT-DILI and UGTs polymorphisms. The statistical analysis strategies refer to previous literature . Allele geneity) . SubgrouThe flowchart for the selection of studies is presented in P = 0.857, P = 0.909, P = 0.393, Among the seven UGT genes, the UGT1A1 gene was the most frequently reported; it was reported in 5 case-control studies , 18–21 iP = 0.009, P = 0.04) [Four SNPs  were reported by two different studies , 20, 21. = 0.04) .P < 0.0001, P < 0.0001, P < 0.0001, P < 0.0001, P < 0.0001, P = 0.007, Three case-control studies with 413 cases and 623 controls focused on the relationships of UGT2B7 polymorphisms with AT-DILI in a Chinese population , 32. FouP = 0.036) [P < 0.001; 308C/A: OR = 3.399, 95% CI: 2.185–5.287, P < 0.001; and rs2070959: OR = 2.342, 95% CI: 1.493–3.675, P < 0.001) [P = 0.006; rs17868324: OR = 2.391, 95% CI: 1.597–3.579, P < 0.001; and rs11692021: OR = 2.383, 95% CI: 1.523–3.729, P < 0.001) [Five other genes in the UGT family with 14 SNPs  were reported in five studies among Korean , 29, 30.= 0.036) . Three SThe present study aimed to verify whether current evidence supports the relationship between UGTs polymorphisms and AT-DILI risk. Our meta-analysis included 12 case-control studies involving 32 SNPs in the seven UGT genes. Based on two original studies, the pooled results indicated that UGT1A1 rs3755319 (heterozygote model) might be associated with AT-DILI risk. In addition, UGT2B7 rs7662029  and rs7439366  were also statistically associated with AT-DILI risk. Therefore, genetic variants in UGT1A1 and UGT2B7 may have relationships with susceptibility to AT-DILI; thus, they have potential for use as biomarkers in the anti-TB treatment population.P = 0.020) [However, the SNP rs4148323, which is the most studied SNP to date, was found to have no significant association with AT-DILI risk under any genetic model. Subgroup analysis also obtained similar negative results in Chinese patients = 0.020) . Other o= 0.020) , 18–20.  . Other = 0.020) . That st= 0.020) . Among p= 0.020) . TherefoP = 0.049) [P = 0.026) [For UGT1A1 rs3755319, in silico analysis indicated that the rs3755319 C allele might induce transcription binding changes and reduce UGT1A1 expression . However= 0.049) , and no = 0.049) . Therefo= 0.049) . A previ= 0.026) . Of courIdentification of a genetic predisposition to AT-DILI is of paramount importance. These meta-analysis results indicated that UGT1A1 rs3755319, UGT2B7 rs7662029, and rs7439366 might be associated with the risk of AT-DILI, which would help to identify susceptible populations for liver injury in patients with anti-TB treatment. If used as a test prior to prescription, genotyping of these genes would prevent potential AT-DILI. However, although various genetic polymorphisms have been identified to be associated with DILI susceptibility, few prospective genetic screening tests have met the threshold for clinical application . The maiThis study was the first to summarize all relevant studies investigating the relationships of UGTs polymorphisms with AT-DILI risk under different genetic models and to perform a meta-analysis of the data reported in those studies. The quality of the included studies was high. Nevertheless, the study had several limitations. First, the number of included studies was small, and the sample size was relatively small for determining genetic association, which made it difficult to draw a robust conclusion. Second, the study subjects were only Chinese and Koreans (limited to Asian countries), which minimized the possibility of discovering meaningful genetic associations. Because fewer than ten studies were included, a publication bias test was not performed. Finally, there existed a high heterogeneity for UGT1A1 rs4148328, and I-squares were larger than 60% under different genetic models. The diagnosis of AT-DILI, causality assessment, and adjustment for covariates were not uniform in those studies, which may be sources of potential heterogeneity.The current meta-analysis indicated that UGT1A1 rs3755319, UGT2B7 rs7662029, and UGT2B7 rs7439366 were significantly associated with AT-DILI risk, and these three SNPs may be used as potential genetic risk markers in anti-TB treatment patients. However, this conclusion should be interpreted with caution due to the low number of studies and the relatively small sample size.


Firearms are frequently the primary cause of death in cases involving homicide, suicide, and unintentional injuries on a global scale . In thiWhile autopsy is widely regarded as the gold standard for post-mortem examination, there is growing recognition of the value of imaging techniques, such as computed tomography, in forensic ballistic investigations. Here, we investigated differences in ballistics findings between conventional autopsy and post-mortem computed tomography (PMCT) specifically in gunshot victims within the head and neck region. By synthesizing the best available evidence, this research aims to contribute to the ongoing discourse surrounding the application of PMCT and its potential implications for informed forensic decisions.This systematic review and meta-analysis adhered to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guideline. A comprehensive search of scientific evidence was conducted using various databases, including PubMed, Web of Science, Scopus, Embase, Google Scholar, and Open Access Theses and Dissertations. The search was carried out on June 7, 2023, employing a structured search strategy outlined in the supplementary file (Supplementary Table 1). Eligible studies had to report at least one of the following ballistic findings in the head and neck region: entrance/exit wounds, wound track, or projectile fragments. We excluded case reports, small case series with fewer than 10 cases, and studies with potential overlapping populations.https://jbi.global/critical-appraisal-tools) was used to assess the risk of bias. The assessment of search results was conducted independently by two reviewers using a two-stage process. Initially, titles and abstracts were reviewed, and potentially relevant studies were identified for a full-text evaluation. Any disagreements between the reviewers were resolved through a consensus-based approach or with the assistance of a third reviewer. Using a standardized data extraction form, we recorded data on authorship, publication year, study country, sample size and characteristics, details of the examinations conducted on gunshot victims, and ballistic findings. The Joanna Briggs Institute Critical Appraisal Checklist for Analytical Cross-Sectional Studies .To ensure a robust and accurate analysis, we followed standard statistical procedures in our study. First, we calculated the prevalence of each ballistic finding in the head and neck region, comparing results from both PMCT and autopsy. Subsequently, we utilized a random-effects model to synthesize prevalence ratios (PR) with 95 % confidence intervals (CI) and visually presented them using forest plots. We assessed the degree of heterogeneity between studies using the IThe initial search yielded 704 references. After screening the titles and abstracts, 31 full-text articles were assessed for eligibility and seven studies  and wound track  between PMCT and autopsy However, PMCT had a 26 % higher prevalence of projectile fragments found than conventional autopsy  .We observed no significant differences in ballistic findings for entrance/exit wounds , Brazil.


Prostate cancer is a slowly progressing cancer. However, it has remained a major medical problem for affected men. Risk factors of prostate cancer include age, race, and prostate cancer family history. Prostate cancer may occur at different frequencies between ethnic populations and countries. Currently, studies on genetic risk factors in prostate cancer aetiology have been increasing. Due to the importance of changes in endothelial nitric oxide synthase in carcinogenesis, we aimed to reveal whether eNOS T786C polymorphism is associated with prostate cancer.Archival samples included in this study were whole blood samples taken from patients who were grouped according to prostate biopsy pathology results  and from healthy participants . DNA was isolated from these whole blood samples and real-time polymerase chain reaction analysis was performed for endothelial nitric oxide synthase T786C polymorphism with LightCycler 480 II. Measured free and total prostate-specific antigen serum levels were evaluated retrospectively.There was a statistical difference between patient-healthy control and control-healthy control groups regarding genotype distributions for eNOS T786C hism. Controls were more likely to have TC and CC genotypes and C alleles than the other two groups.Compared to other groups, the percentage of the eNOS786C allele in the control group was found to be higher. As a result of these data, it can be thought that carrying the allele may be protective against the disease. Prostate Cancer (PCa) is the most commonorgan cancer in males worldwide For the early diagnosis of PCa, serum prostatespecificantigen (PSA) level and digital rectal examination(DRE) combination are recommended asannual screening after the age of 50 Two of the molecules investigated are NitricOxide (NO), which has been shown to be related tothe aetiology of PCa, and nitric oxide synthase (NOS),the enzyme that enables it to be synthesised Since the exact aetiology of PCa is not entirelyunderstood, research on PCa continues. PCa usuallyhas a poor prognosis, and it is thought that the linksbetween tumour development and clinical outcomescan be determined by genetic diversity to determine prognosis This study was approved by the MersinUniversity Clinical Research Ethics Committee datedNovember 25, 2020, numbered 2020/758.One hundred seventeen patients’ samples wereincluded in the study; 48 samples were grouped asthe patient group, 42 as the BPH group, and 27 asthe healthy control group. Archive samples includedin this study were whole blood samples taken frompatients who applied to the urology outpatient clinic,had a prostate biopsy and were grouped according topathology results. According to prostate pathologyresults, patients diagnosed with the malignant prostatedisease were included in the patient group (meanage: 62.83±7.96), and those diagnosed with benignprostate diseases were included in the BPH group. Thepatient group were newly diagnosed with PCa, did nottake medication for PCa, were not hospitalised foranother malignancy or chronic disease in the pastyear, and did not have distant metastases. The BPHgroup did not take medication for other diseases andwere not hospitalised for another malignancy or chronicdisease in the past year. In the healthy controlgroup participants (mean age: 64.63±11.25), theprostate biopsy was not performed, and a PSA valuebetween 2–10 µg/L and with no prostate cancerfamily history, no drug use, no chronic disease, andno hospitalisation history.Free PSA (fPSA) and total PSA (T-PSA) serumlevels were measured in Advia Centaur XP  hormone autoanalyser using the directchemiluminometric sandwich immunoassay method,f/T PSA ratio, and fPSA% values were analysed andevaluated retrospectively. The reference range of T-PSAwas 0–4 µg/L.DNA isolation from archive whole blood takenin ethylene diamine tetraacetic acid (EDTA) containingtubes was performed in accordance with themanufacturer’s instructions  and stored at 4°C.Primers used for eNOS T786C polymorphismanalysis were synthesised by standard phosphoramidechemistry ,and all fluorophore-labelled probes were synthesisedby Metabion and purified by reverse-phase HPLC:Forward Primer: 5’-CCACCAGGGCATCA-AGCT-3’Reverse Primer: 5’-CGCAGGTCAGCAGAGAG-ACTA-3’The Vic probe and fam probe were used todetect the T786C mutation. 5 ends of the Vic probewere marked with Yakima Yellow. Fam probe was a13-mer oligonucleotide:5’-Yakima Yellow-CTGGCTGGCTGAC-3’5’-Fam-TCCCTGGCCGGCT-3'PCR conditions for eNOS T786C polymorphismanalysis were as follows:Denaturation for 600 seconds at 95°C;Denature 15 seconds at 95°C;Annealing for 60 seconds at 60°C (40cycles);Extension for 30 seconds at 40°C.eNOS T786C polymorphism was analysed withallelic discrimination with RT-PCR .Analysis of data was carried out using a trial versionof IBM SPSS Statistics Processor .Comparisons of more than 2 groups in terms ofage variable were evaluated with ANOVA.Control of the normal distribution of numericalPSA variables obtained from these individuals wasperformed by Kolmogorov-Smirnov and Shapiro-Wilknormality tests. PSA variables that were not compatible with normal distribution were summarised withmedian percentages (25%–75%) and minimum-maximumstatistics. In comparing the two groups interms of PSA and its derivatives, the non-parametrictwo independent groups Mann-Whitney U test wasused. Comparisons of more than 2 groups in terms ofvariables were evaluated with Kruskal-Wallis statistics.Pearson Chi-square test results were used to controlthe relationship between categorical variables. Hardy-Weinberg (HW) balance test was used to examine thedistribution of alleles by genotypes and the compatibilityof this distribution with expected values. Possiblerisks of genotypes and alleles were determined by calculatingthe ODDS ratio. The statistical significancelevel was determined as p<0.05.Age values in all groups showed a homogeneousdistribution. The ratio of Hypertension, DiabetesMellitus, other diseases, and PCa presence in thefamily in the groups are given in There was a significant difference between thecontrol and patient groups in terms of T-PSA andfPSA variables (p<0.001). There was no significantdifference between the groups in terms of f/T PSAand fPSA% variables (p=0.277). A significant differencewas found between the patient and BPH groupsin terms of T-PSA, f/T PSA and fPSA% variables(p<0.001). However, there was no significant differencebetween the groups in terms of fPSA variable(p=0.353) . When weThe ratios of eNOS T786C polymorphismgenotypes in groups are given in According to this study’s results, the ratios ofeNOS T786C polymorphism genotypes  were close to each other in the patient andBPH groups, but the healthy control group’s ratio washigher than the other groups. As a result of the within-group comparison in all three groups, no significantdifference was detected between eNOS T786Cpolymorphisms genotype distribution and PSA and itsderivatives. There was a statistically significant differencebetween patient-control and BPH-control groupsregarding genotype distributions for eNOST786C polymorphism. Controls were more likely tohave TC, CC genotypes, and C alleles than the othertwo groups. In the third comparison, the patientgroup was more likely to have TC and CC genotypesand C alleles than the BPH group, but there was nosignificant difference.Various studies have shown that eNOS is effectivein cancer-related processes such as angiogenesis,apoptosis, invasion and metastasis. It has been suggestedthat the eNOS T786C polymorphism affectsthe carcinogenesis process by causing mutant alleles,altered eNOS activity and NO concentrations.Although there are studies investigating the relationshipbetween eNOS polymorphisms and cancer risk,the results appear to be conflicting Abedinzadeh et al. In a meta-analysis, ORs from 11 studies associatedwith the eNOS T786C polymorphism were combined.In the subgroup analysis based on ethnicity,high cancer risk was found in Caucasians but not inAsians, and this difference was associated with asmall number of studies conducted on Asians withdifferent ethnic backgrounds. When stratified by cancertype, a significant relationship was found betweenthe eNOS T786C polymorphism and the increasedrisk of PCa It has been suggested that a dose-dependentrelationship exists between NOS expression andcancer response. It has been reported that high concentrationsof NO have an anti-neoblastic effect bycausing apoptosis, and in low concentrations, proangiogenic and pro-cancerous effects predominateby protecting cells from apoptosis. Some studieshave shown that endogenous NO levels were higherin cancerous prostate tissue than in normal tissue.Conversely, in another study, NOS expression decreasedin BPH developing in the transitional zone ofthe prostate, and consequently, there was a decreasein NO production Since the eNOS T786C polymorphism is locatedin the gene’s promoter region, it is important tobind RNA polymerase to this region and the gene’stranscription. Polymorphism in this region was thoughtto trigger the development of PCa by causingthe increase or decrease of NO synthesis, althoughthe underlying mechanisms are not fully known When we evaluated the results of this study, wecould not find a relationship between eNOS T786Cpolymorphism genotypes and PSA, PSA derivatives,and PCa. However, the percentage of the C allele inthe control group was found to be higher in genotypeand allele comparisons. These data show that the riskof prostate cancer is low in those carrying the C allele,suggesting that carrying the C allele may be protectiveagainst the disease. Considering that no tumour marker can clearly distinguish between benign andmalignant, the high rate of the C allele compared tothe control group is a significant result.Although several case-control studies have beenconducted to assess the role of eNOS gene polymorphismsin susceptibility to PCa in different populations,conflicting results have been reported due to the relativelysmall sample size of the individual studies andthe effects of the sample group. Apart from this, racialdifferences in polymorphism studies significantlyaffect the results.The selected group may have caused the datadifferences in the literature, the number of participants,the use of different samples, and the differencein the participants selected as controls , the average age of the participants.We believe that the inconsistencies in the literaturecan be overcome by studies involving moreparticipants, perhaps different racial subgroups, subdividingwhen necessary, and simultaneously measuringNO levels.PCa, prostate cancer;
PSA, prostate-specificantigen;
DRE, digital rectal examination;
TRUS, transrectal ultrasound;
NO, nitric oxide;
NOS, nitric oxide synthase;
nNOS, NOS1,neuronal NOS;
iNOS, NOS2, inducible NOS;
eNOS, NOS3, endothelialNOS;
BPH, benign prostatic hyperplasia;
fPSA, free PSA;
TPSA,total PSA;
fPSA%, percentage of free PSA;
f/T PSA, free/totalPSA ratio;
EDTA, ethylene diamine tetraacetic acid;
HW, Hardy-Weinberg Balance Test;
HT, hypertension;
DM, diabetes mellitus.We thank H. Didem OVLAfor the statistical analysis of our research.This study was approved bythe Mersin University Clinical Research EthicsCommittee dated November 25, 2020, numbered2020/758.SB planned the work, performedthe analysis, evaluated the results, and wrotethis article; SA planned the work, made a selection ofsamples, evaluated the results, and helped to writethis manuscript; LT participated in its design andcoordination, and helped to draft the manuscript; MBmade a selection of samples, helped to evaluate theresults of the analysis, and helped to draft the manuscript.All authors read and approved the final manuscript.All expenses of the study were covered by theresearchers.All the authors declare that they have no conflictof interest in this work.


Hundreds of randomised controlled trials and dozens of meta-analyses have examined psychotherapies for depression—yet not all points in the same direction. Are these discrepancies a result of specific meta-analytical decisions or do most analytical strategies reaching the same conclusion?We aim to solve these discrepancies by conducting a multiverse meta-analysis containing all possible meta-analyses, using all statistical methods.p-uniform and PET-PEESE (precision-effect test and precision-effect estimate with SE) meta-analysis models. This study was preregistered (https://doi.org/10.1136/bmjopen-2021-050197).We searched four bibliographical databases , including studies published until 1 January 2022. We included all randomised controlled trials comparing psychotherapies with control conditions without restricting the type of psychotherapy, target group, intervention format, control condition and diagnosis. We defined all possible meta-analyses emerging from combinations of these inclusion criteria and estimated the resulting pooled effect sizes with fixed-effect, random-effects, 3-level, robust variance estimation, gmean=0.56, a medium effect size, and ranged from g=−0.66 to 2.51. In total, 90% of these meta-analyses reached a clinically relevant magnitude.A total of 21 563 records were screened, and 3584 full texts were retrieved; 415 studies met our inclusion criteria containing 1206 effect sizes and 71 454 participants. Based on all possible combinations between inclusion criteria and meta-analytical methods, we calculated 4281 meta-analyses. The average summary effect size for these meta-analyses was Hedges’ The multiverse meta-analysis revealed the overall robustness of the effectiveness of psychotherapies for depression. Notably, meta-analyses that included studies with a high risk of bias, compared the intervention with wait-list control groups, and not correcting for publication bias produced larger effect sizes. To this day, more than 80 meta-analyses have examined the efficacy of psychotherapies for depression. However, recent meta-research projects questioned whether the effect sizes from those published meta-analyses were inflated and discrepancies between these meta-analyses emerged.In a so-called multiverse meta-analysis, we calculated over 4000 meta-analyses—most of them (90%) produced small but clinically relevant effect sizes. Our findings suggest that psychotherapies for depression are generally effective, but the specific type, format and other factors can affect the magnitude of the treatment effect slightly. However, meta-analyses that (1) restrict their control group to wait-list control groups, (2) do not exclude high risk of bias studies and (3) do not correct for publication bias are likely to produce larger effect sizes.The effect size reported in any given meta-analysis on treatment efficacy for depression depends less on the type of psychotherapy, treatment format, diagnosis or target group but rather on the comparison with wait-list control groups, not excluding the high risk of bias studies, or correcting for publication bias. In general, future meta-analyses that diverge from small-to-medium summary effect size estimates may be indicative of extreme data analytical decisions and may therefore not contribute additional substantive knowledge. However, there are certain circumstances where higher treatment effects might be observed, such as when a future meta-analysis examines a specific subgroup of patients or a new type of therapy not previously studied.Over the last four decades, more than 80 meta-analyses have examined the efficacy of psychotherapies for depression.Some of the discrepancies in findings may be the result of publication bias leading to an overestimation of the effectiveness of psychotherapy or may be due to variations in inclusion criteria, such as the inclusion of low-quality studies or studies comparing interventions with wait-list control groups only.Although conventional meta-analyses exist on some of these specific aspects, a comprehensive bird’s-eye view of all meta-analyses for depression treatment research is missing. Ideally, this birds-eye view does not only include all published meta-analyses but also all possible meta-analyses based on defensible and reasonable analytical choices. To provide such an overview and fill substantive knowledge gaps, we conducted a so-called multiverse meta-analysis and calculated all possible meta-analyses on the efficacy of psychotherapies for depression in a single analysis. It can (1) integrate multiple meta-analyses like an umbrella review, (2) enable us to identify knowledge gaps and (3) investigate how flexibility in data selection and analysis might affect the overall interpretation of results. In doing so, this new approach can help solve diverging claims on the efficacy of psychotherapies once and for all. We replicated most meta-analyses that have ever been conducted in research on psychotherapy for depression and created additional evidence by conducting thousands of meta-analyses that were still missing in the literature.Due to the sheer number of published meta-analyses and primary studies on these differences between psychotherapies, we aimed to summarise, integrate and visualise the entire evidence. As a result, we can highlight robustness—or lack thereof—by inspecting all possible meta-analyses to help resolve conflicting meta-analyses and contested evidence, alleviate the associated adverse effects of these phenomena on research progress and provide a birds-eye perspective of the entire field.https://doi.org/10.1136/bmjopen-2021-050197).We searched four major bibliographic sources .g) for postintervention comparisons between psychotherapy and control conditions based on continuous outcome data provided in the primary studies. If only change score or dichotomous outcome data were reported, we converted these data into Hedges’ g with the metapsyTools R package.Our study protocol outlines the analyses in more detail  and statistical methods investigating psychotherapies targeting depression that could possibly be conducted.Which factors—asking which data to meta-analyse—and one How factors—asking how to meta-analyse the data. Based on these Which factors, meta-analyses could include studies investigating different target groups that received different types of psychotherapies in different formats and assessments/diagnoses of depression. These studies could have different risks of biases’ ratings and could be compared with different control groups (see Which factors). Based on our How factor, we used six different meta-analytical methods to analyse the data: random-effects, fixed-effect, 3-level, robust variance estimation (RVE),p-uniform* meta-analysis models.We specified seven the so-called r=0.5) and introduce a unit-of-analysis problem.Some primary studies reported multiple effect sizes per study, that is, when multiple instruments were used to measure depression or when a study consisted of multiple interventions or control groups. These nested effect sizes are not independent as they are correlated , reporting bias  or clinical heterogeneity . This phenomenon, which is common in many scientific fields, can cause inflated effect sizes and therefore an overestimation of treatment effectiveness.small-study effects in meta-analyses.PET-PEESE is a regression-based method for addressing the issue of PET-PEESE aims to correct for this bias and provides more accurate estimates of effect sizes.P-uniform* is a selection model approach that uses a random-effects model as its effect size model.P-uniform* works by treating the primary studies’ effect sizes differently based on whether they are statistically significant or not. This method can be considered a selection model approach with a single cut-off value that determines whether an effect size is considered statistically significant.g=0.24, which was suggested as the minimal important difference for interventions targeting major depressive disorder.The results of the multiverse meta-analysis were depicted in a descriptive specification curve plot, which is a graph that shows the results of all conducted meta-analyses, represented by points on the graph with 95% CIs. The summary effects are plotted in order of magnitude, from lower to higher, and connected by a solid line—the specification curve. To evaluate the number of meta-analyses that exceed relevant magnitudes, the plot focuses on two thresholds: meta-analyses that do not include 0 in their 95% CI  and meta-analyses that exceed a clinically relevant effect size of Hedges’ In a second step, we evaluated if the findings of the descriptive specification curve  are likely to be true or if they could be due to chance. This so-called inferential specification curve analysis involves simulating new random-effect sizes for each primary study under the assumption that the null hypothesis, or the assumption that there is no psychological treatment effect on depression, is true.These new data sets are created by drawing random values for the effect sizes from a normal distribution with mean zero and an SD that takes into account both the variance of the original effect size estimate for each study and a measure of between-study heterogeneity .We then applied a new descriptive specification curve analysis under three scenarios: a fixed-effect scenario with no heterogeneity, a scenario with heterogeneity equal to the random-effects model of the 415 studies and a scenario with the upper 95% CI estimate of τ from the random-effects model. By repeating this process 1000 times, we were able to identify the lower and upper limits of the resulting 1000 bootstrapped specification curves or their 2.5% and 97.5% quantiles. These limits, or quantiles, represent the range within which we would expect 95% of all meta-analyses to fall if the true underlying effect were a null effect. If the observed treatment effect falls outside of these limits, it is considered to be a deviation from the null hypothesis and is considered to be a likely true effect.415=8.46×10124.Additionally, we conducted a GOSH plot  to identify the overall range of possible summary effect sizes for meta-analyses and visualise the heterogeneity of effect sizes.one possible meta-analysis—including the broadest inclusion of all Which factors and the 3-level How factor—out of the entire multiverse of meta-analyses. The amount of heterogeneity was estimated using the restricted maximum likelihood estimator. In addition to the estimate of To create a reference point for exploring heterogeneity in the data, we additionally fitted a 3-level model to the entire data set. This resulting summary effect size represented esk=1206 effect sizes from studiesk=415 studies were included. The sample sizes of the included primary studies ranged from n=4 to 1156, meanN=103, medianN=67. The total sample size of all included samples from all primary studies was totalN=71 454. See We screened 21 563 titles and examined 3584 full-text papers, of which 415 were included. The PRISMA  flow chart depicts the study inclusion process in Most included primary studies investigated the efficacy of psychotherapies for depression in adults (37%) or general medical populations (23%). Most studies were either conducted in Europe (37%) or in Northern America (36%). Most studies used CBT-based interventions (69%) and were primarily delivered in individual therapy format (35%). Care-as-usual was the most common type of control condition (49%), and depressive disorder was diagnosed by a clinician in 49% of studies. Only 35% of studies were rated with a low risk of bias. See g=−0.66 to 2.51. Half of those effect sizes were in the IQR of Hedges’ g=0.42 to 0.71, representing small-to-medium effect sizes. In total, 97% of the effect sizes were greater than 0, and 84% of these had 95% CIs that did not include 0 . In total, 90% reached a clinically relevant magnitude of Hedges’ g >0.24, and 68% of the summary effect sizes had 95% CIs above the clinically relevant cut-off.Our multiverse meta-analysis produced 4281 unique meta-analyses, with effect sizes ranging from Hedges’ The overall pattern of the descriptive specification curve indicates that larger meta-analyses, including more primary studies, had medium-to-large effect sizes and were close to the median-estimated effect size of the multiverse see . More exWhich and How factors produced—on average—systematically different summary effect size estimates compared with others. In the following, we descriptively summarise the most important results. It is important to note that this examination of these differences is based on descriptive analysis rather than a formal statistical comparison. For a more detailed breakdown of each Which factor, see Several g=0.82, 95% CI  from k=54 included meta-analyses, produced larger effect size estimates than meta-analyses on adults, mean g=0.51, 95% CI  with k=1076.Meta-analyses including only student populations, mean g=0.76, 95% CI  with k=536, produced larger effect size estimates than meta-analyses delivered as guided self-help interventions, mean g=0.50, 95% CI  with k=586.Meta-analyses including studies delivered in a group format, mean g=0.60, 95% CI  with k=1620, produced larger effect size estimates than meta-analyses focusing on non-CBT treatments, mean g=0.48, 95% CI  with k=733.Meta-analyses focusing only on CBT-based treatments, mean g=0.66, 95% CI  with k=836, produced larger effect size estimates than treatments compared with care-as-usual, mean g=0.52, 95% CI  with k=1194.Meta-analyses that included samples compared with a wait-list control group, mean g=0.61, 95% CI  with k=2413 included samples, produced larger effect size estimates than meta-analyses including only low risk of bias studies, mean g=0.45, 95% CI  with k=1034.Meta-analyses that excluded high risk of bias studies, mean g=0.56 and 0.57, respectively.Meta-analyses that included studies in which depression was diagnosed by a clinician or was self-reported produced similar results, mean g=0.66, 95% CI  with k=963, produced larger effect size estimates than meta-analyses analysed with PET-PEESE, mean g=0.18, 95% CI  with k=591.Meta-analyses analysed with 3-level models, mean How factors rather than different Which factors. Of the 688 meta-analyses that included a zero in their 95% CI, the PET-PEESE method accounted for 408, the RVE estimation method for 143 (as it tends to produce larger CIs), p-uniform for 68 and 3-level modelling for 59. The fixed-effect  and REML models  barely produced any statistically non-significant meta-analyses. No other systematic differences were observed among those meta-analyses.We further investigated the studies in which there was no strong evidence that psychotherapies were effective indicated by including zero in the 95% CI. On closer inspection, we found that the observed null effects were largely caused by using different g=0). This was true for a range of scenarios, from simulating no heterogeneity (τ=0) to the identified heterogeneity in the random-effects model of all 415 studies (τ=0.53) and the respective 95% CI upper limit of τ=0.71. In fact, 98% of the meta-analyses were outside the expected range under the scenario of no effect, suggesting that the meta-analyses found in the multiverse meta-analysis were significantly different from a null effect. In addition, we conducted a similar analysis for a scenario in which the simulated studies had a clinically relevant effect size of Hedges’ g=0.24, rather than g=0. We found that, consistent with our previous analysis, most treatments were more effective than the simulated effect sizes in this scenario , p<0.001). The estimated variance components were τ2Level 3=0.364 and τ2Level 2=0.004. Overall, I2Level 3=86% of the total variation can be attributed to between-study and I2Level 2=1% to within-study heterogeneity. We found that the 3-level model provided a significantly better fit compared with a 2-level model with level 3 heterogeneity constrained to zero . The overall I2 value, indicating how much of the total variance can be attributed to the total amount of heterogeneity, is very large, with approximately 87% of the total variance attributable to heterogeneity.We chose to include a conventional 3-level meta-analysis, including all 1206 effect sizes from all 415 primary studies, as an exemplary meta-analytical specification out of the 4281 meta-analyses. The pooled Hedges’ We investigated the efficacy of psychotherapies for depression by simultaneously analysing 4281 meta-analyses based on all reasonable combinations of inclusion criteria and meta-analytical methods. We investigated the influence of different treatment groups, types of psychotherapy, treatment formats, control groups, diagnoses for depression, risk of bias assessments and different meta-analytical methods. We found that most meta-analyses produced small-to-medium, but clinically relevant effect sizes, suggesting the overall robustness of psychotherapies for depression. These results are supported by our descriptive and inferential multiverse meta-analysis and the combinatorial meta-analytical approach. Notably, specific patterns emerged in the descriptive specification curve analysis.Meta-analyses that compared interventions with wait-list control groups had larger effect sizes compared with those including care-as-usual or other control conditions, such as attention placebo. This aligns with previous research suggesting that effect sizes obtained in RCTs should be interpreted differently depending on the used control condition and warning especially against lumping control conditions into one comparison group for network meta-analyses.Our findings indicate that the effect sizes found in depression research might be inflated, even when high risk of bias studies are removed. To support this conclusion, we compared the results of meta-analyses that excluded high risk of bias studies with meta-analyses that did not exclude them. We found that excluding high risk of bias studies did not result in a reduction in effect size estimates. In fact, the effect sizes were very similar to those observed when all studies, regardless of study quality, were included in the meta-analysis. However, when we only included studies with a low risk of bias in our meta-analyses, we did observe a reduction in effect size estimates. This finding is in line with earlier research that has also suggested that effect sizes in depression research may be inflated, as including only low risk of bias studies yields substantially smaller effect size estimates.Our analysis of publication bias  in psychotherapy research on depression showed that one of the methods used to correct publication bias resulted in substantially smaller effect size estimates compared with conventional methods that did not correct for publication bias. Even though the PET-PEESE estimator may have overcorrected for biases in our data, resulting in negative effect sizes, these negative effect sizes indicate that the meta-analyses under study had no effect when correcting for small-sample effects.In addition, we found that meta-analyses focused on student populations and interventions delivered in a group format had higher effect size estimates. This finding is consistent with previous meta-analyses that found interventions to be more effective in young adults compared with middle-aged adults,g>3). These enormous effect sizes might distort meta-analyses with only a few included studies towards more extreme summary effect sizes. For this reason, we included only meta-analyses with more than 10 studies in our multiverse meta-analysis to avoid such extreme meta-analyses. We assessed the influence of several cut-offs in Which factor categories to ensure that both computations and visualisations remained feasible. For instance, we initially planned to investigate each therapy type individually, yet we had to create broader categories and combined CBT-based and non-CBT-based therapy approaches together. These merges decreased the level of detail of our analyses, but at the same time ensured the interpretability, as a visualisation of over 40 000 meta-analyses (resulting from not merging different therapy types) was simply not possible. Third, the methods we used for correcting for publication bias have some limitations in performing in environments with low sample size and high heterogeneity—as was the case with the body of evidence in this multiverse meta-analysis. This might have caused the PET-PEESE method to underestimate and p-uniform to overestimate the effect size. Overall, the presence of publication bias and its inflating influence on effect size estimates remain highly likely and have been reported in other publications as well.Our multiverse meta-analysis was able to demonstrate that the inclusion of high and medium risk of bias studies, the comparison with wait-list control groups and models not accounting for publication bias yielded larger—and potentially inflated—effect size estimates for the efficacy of psychotherapies for depression. At the same time, we were able to demonstrate that even after considering this inflation, the effectiveness of psychotherapies remains clinically significant. Despite these strengths of our study, several limitations need to be acknowledged. First, our database contains several studies with unreasonably large effect sizes  network meta-analyses, as well as longitudinal approaches, are needed to investigate these more relevant and critical issues for the individual patient. Finally, this study provides the most exhaustive overview of psychological depression research that is available and possible today. It can guide future research as knowledge gaps were closed and is a valuable source for policy-makers to inform evidence-based decision-making.In summary, our multiverse meta-analysis completes the evidence that psychotherapies are effective for treating depression in a wide range of patient populations. Because we evaluated the entire multiverse of defensible combinations of inclusion criteria and statistical methods on depression research, our results suggest that this line of research and the debate about whether treatment is effective can now end once and for all. Future research can and should be less concerned with


This systematic review and meta-analysis summarize the evidence for the association between Helicobacter pylori infection and Primary Open-Angle Glaucoma.Eligible studies reporting an association between H. pylori infection and Glaucoma were identified through an extensive search of the Excerpta Medica (EMBASE), Web of Science, Scopus, and PubMed databases and an assessment of the reference list of the top articles until October 2022. Analysis was performed with random effects model using Stata 16.Twenty-four studies were included in the systematic review. This study involved 1602 glaucoma patients and 2800 control individuals. The combined RRs of cohort studies and overall combined ORs of case-control studies showed a significant correlation between H. pylori infection and Glaucoma. Subgroup analysis showed that glaucoma patients had a higher risk of having H. pylori infection if they were residents of Europe countries  and , if they had POAG type  and , if their diagnostic method of HP was histology  and , and if they were over 60 years old  and .The results of this meta-analysis suggest a statistically significant association between Helicobacter pylori infection and Primary Open-Angle Glaucoma. Helicobacter pylori (HP) is a spiral-shaped and gram-negative Micro-organism that affects the epithelial mucosa of the stomach , 2. It iGlaucoma is the acquired loss of retinal ganglion cells (RCG) and axons within the optic nerve or optic neuropathy . GlaucomThe hypothesis of the correlation between H. pylori infection and glaucoma originated from the role of H. pylori causing the release of various pro-inflammatory and vasoactive substances, arteriosclerosis-induced increased platelet activation and aggregation, and impact on the trabecular meshwork cell apoptotic process. Cross-reactivity between antibodies to Helicobacter pylori and the ciliary epithelium has also been speculated –22.As studies evaluating potential associations between H. pylori infection and POAG have reported positive , 23–38 oThis study aimed to address the Association between Helicobacter pylori Infection and POAG through systematic review and meta-analysis based on the PRISMA guidelines for systematic reviews and meta-analysis of cohort and case studies . A systeIn the current meta-analysis, the following inclusion and exclusion criteria were required to be met. The inclusion criteria were: cohorts or case-control studies evaluating the association between H. pylori infection and Glaucoma and their results using the indicators measuring the association, such as OR, RR, and HR with a 95% confidence interval. The exclusion criteria were duplicate data, Cross-sectional studies, editorials, letters, review articles, case reports, systematic reviews, and meta-analyses, intervention studies (RCT), and articles not reporting the outcomes of interest.At the end of the search, all screened studies were uploaded to Endnote version 9. After identifying and removing duplicates, studies still needed to be published and were in the peer-reviewed phase were excluded. After that, the full texts of the screened articles were reviewed, and the ones which met the desired criteria were selected for meta-analysis. If multiple types of glaucoma were reported in an article, each kind combined with the control group was extracted as an independent dataset.After selecting the studies, the required data were extracted by the two independent researchers (EEA and YM) and recorded in Excel. After examining the discrepancies and reaching a single result, the OR and RR reported with a 95% confidence interval and related to the association between H. pylori and POAG were extracted from the studies. Then, this information was removed, including the first author’s name, year of publication, Type of study, country of origin, Host Journal, sample size, Glaucoma Cases, Controls, Mean Age (Years), Glaucoma Subtype, H. pylori Infection Diagnosis, NOS Score and Relationship between H. pylori and Glaucoma .The quality of the studies in this meta-analysis was assessed using the Newcastle Ottawa Scale (NOS) , 47. On In this systematic review, seven cohort and seventeen case-control studies were included in the analysis. The researchers have decided to perform the final analysis on both cohort and case-control studies since they are methodologically different, and their indicators of measuring the association are various. Analyzes related to these studies were performed by Stata software version 16. First, the logarithm and the standard deviation (SD) of the relative risk (RR) logarithm were calculated using a combination of seven cohort studies. Second, the logarithm and the standard deviation of the logarithm of Odd’s Ratio (OR) were computed using a combination of seventeen case-control studies. The model of fixed effects was used to estimate the pooled risk ratio. Also, subgroup analyses were performed to identify the primary source of heterogeneity by considering the Continents where the study was conducted, the type of Glaucoma, the Diagnostic Method of HP, and the participants’ age. I square index and Cochrane Q test were used to determine heterogeneity. Egger and funnel plot tests were used to evaluate the publication bias. The significance level was considered below 0.05 in this study.After completing the search in international databases, 826 articles were retrieved. First, duplicates (541) were removed, then 285 articles were entered into the screening stage according to the title. One hundred sixty-six articles were removed at this step, and 119 papers were evaluated based on their abstracts, considering the inclusion and exclusion criteria. Then, forty-eight pieces remained and entered the screening phase based on their full texts. Of these articles, twenty studies were excluded due to the non-related outcomes to the present study and eight cases because of different statistical populations. Finally, twenty studies related to the subject and purpose of the research had the necessary conditions to be entered into the study. After a manual search, four articles were added to the final studies , 37, 48.After screening, twenty-four articles were finally selected, of which seven articles were cohorts , and the highest effect belonged to the study of Kountouras et al. with the risk ratio of 3.27 . Finally, after combining these cohort studies, the pooled risk ratio of association between H. pylori and Glaucoma was 1.59 . The heterogeneity in this analysis was low and equal to 12.17%, with a significance level of 0.001 , and the highest effect belonged to the study of Xinjifu et al. with the odd’s ratio of 15.84 . Finally, after combining these case-control studies, the pooled odds ratio of association between H. pylori and Glaucoma was 2.87 . The heterogeneity in this analysis was equal to 72.13%, with a significance level of 0.001  was observed in the combination of studies included in our analysis. However, the method of cutting and filling the collected effect size did not change the results, which showed that the study results were not affected by publication bias Figs.  and 6.The results of subgroup analysis based on the continent where the study was conducted, Type of Glaucoma, Diagnostic Method of HP, and Age are shown in Table Subgroup analysis based on the geographical location in the present meta-analysis showed that glaucoma patients had a higher risk of having H. pylori infection if they were residents of Europe countries  and  (Table ).Subgroup analysis based on the type of Glaucoma in the present meta-analysis showed that glaucoma patients had a higher risk of having H. pylori infection if they had POAG type.  And  (Table 2).Subgroup analysis based on the diagnostic method of HP in the present meta-analysis showed that glaucoma patients had a higher risk of having H. pylori infection if their diagnostic method of HP was histology.  And   and  (Table In recent decades, HP infection has been considered a critical risk factor for glaucoma. However, the study sample size is limited, and the studies have conflicting results. Therefore, we conducted this meta-analysis to determine the relationship between Helicobacter pylori and primary open-angle glaucoma.In the present meta-analysis, twenty-four studies were reviewed and included. These studies included case-control and cohort studies, and each study was separately analyzed. The results of this meta-analysis, both in case-control studies and cohort studies, showed the relationship between HP and Glaucoma. In addition to the overall association of Glaucoma with HP in this study, subgroup analysis showed the association of HP infection with POAG, NTG, and PXFG. In addition, subgroup analysis based on geographical location in the present meta-analysis showed that glaucoma patients are at higher risk of HP infection if they live in European countries. Also, in this meta-analysis, the age of Glaucoma patients in selected primary studies was divided into groups less than or equal to 60 years and more than 60 years. The analysis showed that glaucoma over 60 increases the risk of HP infection.Glaucoma is the second leading cause of blindness worldwide after cataracts  and the Another point is that studies have proven that glaucoma patients have a common genetic factor that makes them more susceptible to HP infection . In addiAs mentioned, there are various studies on the relationship between HP and Glaucoma and its types, including POAG. For example, some studies have a confirming role in this connection and investigate the condition of Glaucoma with the treatment of Helicobacter pylori. In a study by Shahram Ala et al., it was stated that HP eradication treatment might positively affect glaucoma management. The study results showed that intraocular pressure decreased significantly after HP eradication treatment in the intervention group. At the same time, this was not the case in the control group that was not treated with HP . The samOther studies have documented a higher prevalence of HP in POAG. These studies concluded that HP infection is much more common in patients with primary open-angle glaucoma , 40. In The current study had strong points compared to previous meta-analysis studies, including that the last field meta-analysis  was condIn addition to these strengths, the Egger test results showed an evident and significant publication bias in the combination of studies included in our analysis. However, the cutting and filling method did not show publication bias. Nevertheless, despite a comprehensive search of all relevant articles, publication bias is still unavoidable.The presence of our important confounding variables between glaucoma and Helicobacter pylori may help health policymakers, and clinical professionals make decisions. One of the limitations of this meta-analysis is the lack of subgroup analysis based on confounding variables based on the presence of chronic diseases such as diabetes, HTN, and … or parameters like BMI. This is the need for more reporting of the desired relationship in the present meta-analysis based on these variables in selected primary studies. Future studies should be conducted considering these variables with a large sample size.The present study provided strong evidence regarding the relationship between HP and Glaucoma. Based on the results of this study, HP is considered a risk factor for Glaucoma, as well as its types, including POAG, NTG, and PXFG. Many factors can cause Glaucoma, and simply by eradicating Helicobacter pylori, we can’t expect that glaucoma will disappear; only perhaps its incidence will decrease. So, further studies and sufficient evidence on a larger scale are needed to show that the elimination of Helicobacter pylori infection positively impacts glaucoma parameters.