Inconsistent findings existed on the correlation of collagen type V α1 (COL5A1) gene polymorphisms and musculoskeletal soft tissue injuries (MSTIs). The purpose of this study was to collect and combine the current evidences by a meta-analysis approach.Six online databases were searched up to August, 2021. The methodological quality of each individual study was evaluated based upon Newcastle–Ottawa Scale (NOS). The strength of the effect size was presented by odds ratio (OR) with 95% confidence interval (95%CI) in five genetic models. The data were analyzed using Review Manager 5.3.Twenty-one studies were eligible to this meta-analysis. The study quality was deemed fair to excellent according to NOS. In the overall analyses, the merged data suggested that rs12722, rs71746744, and rs3196378 polymorphisms were correlated to an increased susceptibility to MSTIs. But the association was not established in rs13946 or rs11103544 polymorphism. For rs12722 polymorphism, stratified analyses by injury type and ethnicity identified the association mainly existed in ligament injury and among Caucasian population. For rs13946 polymorphism, subgroup analysis suggested the association existed in tendon and ligament injuries.This study supports that rs12722 is associated with an elevated susceptibility to ligament injury, especially in the Caucasian population. Rs13946 polymorphism appears to increase the risk to tendon and ligament injuries. Rs71746744 and rs3196378 polymorphisms have a tendency to confer an elevated risk to MSTIs. However, no relevance is found between rs11103544 polymorphism and MSTIs. Musculoskeletal soft tissues, including tendon, ligament, and muscle, are commonly injured as a result of participation in competitive and recreational activities. It has been estimated that over 100 million musculoskeletal soft tissue injuries (MSTIs) occur annually around the world , 3. MSTIA fully recognition on the etiology of MSTIs is of great important to prevent these injuries. Nevertheless, the etiology of MSTIs is multifactorial and its pathogenesis remains largely undefined. Both genetic and non-genetic risk factors have been reported to dispose an individual to MSTIs , 5. Non-The COL5A1 gene codes for the α1 chain of type V collagen. Despite type V collagen presents in a smaller amount than other fibrillar collagens, it exerts a crucial role in fibril assembly and inhibition of lateral fibril growth, leading to fewer collagen I fibrils with increased diameters in tendons and ligaments . LiteratMokone et al.  first reAn exhaustive literature search of PubMed, Web of Science, EMBASE, Cochrane Library, CNKI, and Wanfang databases was performed to look for studies that reported the association of COL5A1 gene polymorphisms and MSTIs. The terms for literature search included “COL5A1”, “tendon rupture”, “tendon injury”, “ligament injury”, “muscle injury”, “soft tissue injury”, “tennis elbow”, “polymorphism”, “variant”, and “mutant”. The references of eligible articles were also screened for potentially relevant studies. No restriction was set on language or publication date. For non-English and non-Chinese literature, they were translated into English paper by native speaker. The final systematic search was conducted on August, 2021. If necessary, the corresponding author of original articles was contacted for additional information.The eligible studies should satisfy the following criteria: (1) studies on the association of COL5A1 polymorphisms and MSTIs; (2) cases were confirmed by clinical evaluation and/or other complementary examination; (3) controls were healthy individuals without MSTIs; (4) data were full to evaluate the odds ratios (ORs) and 95% confidence intervals (95%CI).Correspondingly, the exclusion criteria were as follows: (1) Duplicate data were involved in the studies; (2) conference abstracts, reviews, editorials, or case reports. If multiple studies reported overlapping data, the one with the largest sample size was selected.The assessment of study quality was also performed by two authors (RG and ZJ) individually according to the Newcastl-Ottawa Scale (NOS) , which iRelevant data were abstracted from qualified studies independently by two investigators (RG and ZJ). The data were first author, publication year, country, ethnicity, gender, study design, diagnosis, genotype distribution of each polymorphism in both groups, result of HWE test . Any divI2 test. When significant heterogeneity was achieved , the data was merged with the random-effects model. Otherwise, the data were combined with the fixed-effects model. Based on ethnicity  and diagnosis , subgroup analyses were performed.OR and 95%CI were estimated to evaluate the strength of the association. It was assumed that “V” and “v” represented the mutant allele and the wild allele respectively. Therefore, the genotypes could be represented by “VV”, “Vv”, and “vv”. The pooled effect size was calculated respectively for allele contrast of V versus v, homozygote contrast of VV versus vv, heterozygote contrast of Vv vs. vv, dominant contrast of VV + Vv vs. vv, and recessive contrast of VV vs. Vv + vv. The intra-study heterogeneity was evaluated using Q-statistical test and Sensitivity analysis was performed by sequentially ignoring a single study at a time, which could judge the influence of an individual dataset on the aggregated outcomes. The potential publishing bias was estimated by funnel plots. The data were analyzed by RevMan 5.3 software.https://pubs.broadinstitute.org/mammals/haploreg/Haploreg.php) was used to predict the function and interplay of COL5A1 polymorphic sites. String online server was used to examine the network of gene–gene interaction for COL5A1 gene (https://string-db.org/).Bioinformatics database of HaploReg 4.1 , heterozygote model , and dominant model .Eighteen studies , 38, 39 Stratified analyses by injury type suggested that rs12722 polymorphism was associated with an increased susceptibility to ligament injury under five genetic models. But the association was not found in tendon injury or muscle injury. When stratified by ethnicity, the combined outcomes indicated that s12722 polymorphism was significant associated with MSTIs in Caucasians but not Asians.Seven studies , 35, 38 P < 0.01; TT + TC vs. CC, OR = 2.28, 95%CI 1.23–4.23, P = 0.009) and ligament injury . Because only one study was conducted in Asian, stratified analysis by ethnicity was only carried out in Caucasian. The merged data indicated a null association between rs13946 polymorphism and MSTIs in Caucasians.Subgroup analyses by injury type suggested that rs13946 was significantly associated with an elevated susceptibility to tendon injury .Two studies , 37 withP = 0.02; AA vs. CC, OR = 1.46, 95%CI 1.05–2.03, P = 0.03; AA + AC vs. CC, OR = 1.45, 95%CI 1.111–1.88, P = 0.006).Three studies , 32, 37 After excluding the studies out of HWE, the OR and 95%CI did not reverse. With sequential removal of each study, the pooled OR and 95%CI of the rest studies did not change significantly, indicating the results were stable. The funnel diagrams did not show obvious sign of dissymmetry, suggesting no significant publication bias Fig. .Fig. 3FuThe results from HaploReg indicated that rs12722 was in linkage disequilibrium with rs3196378, and rs13946 was in linkage disequilibrium with several other polymorphic sites  than in Asians . Therefore, the inconsistent outcomes between Asians and Caucasians may be attributable to differences in genetic background.Based upon the current findings, future works should be focus on rs12722 and rs13946 polymorphisms. As each individual has a unique genetic profile, genetic screening tools might be designed to identify individuals predisposed to MSTIs, thus enabling implementation of preventive strategies for them. Correspondingly, taking preventive measures might reduce the incidence of MSTIs and its cost . While, Collagen is best known as the principal tensile element of connective tissues like tendons, ligaments, and cartilage . Type V The five studied loci were located in the 3’-UTR of COL5A1 gene. Despite 3’-UTR has a noncoding character, mutations within this region may modify the secondary structure of mRNA and protein features . Indeed,To investigate the interaction effects between polymorphic sites, functional predictive analysis was performed. The results from HaploReg indicated that rs12722 was in linkage disequilibrium with rs3196378, and rs13946 was in linkage disequilibrium with several other polymorphic sites (Fig. Of note, Lv et al.  had publHowever, several potential drawbacks could not be overcome in this study. First, although subgroup analysis was performed, the heterogeneity in some contrasts still could not be well addressed.The heterogeneity might be explained by diversity of injury types, differences in sequencing methods, variance of ethnic origins, and differences in the selection of participants. Heterogeneity should be considered when interpreting the findings, and future studies should be focused on more homogenous groups of patients. Second, the number of studies of rs13946, rs11103544, rs71746744 and rs3196378 polymorphisms was small. The statistical power might not be strong enough to explore the relationship of the four polymorphisms and MSTIs. Third, clinical heterogeneity, such as age, sex, lifestyle, mechanism of injury, physical or occupational activity, and other potential confounding factors, could not be managed, which might distort the outcomes. Fourth, because the ethnicity subgroup analyses were restricted to European, Asian and Brazilian populations, the results are only applicable to such ancestry groups. Fifth, several of the included studies were out of HWE, which could be caused by population stratification, genotyping errors, and selection bias in the recruitment of controls . Last, bTaken together, the current meta-analysis supports that rs12722 is associated with an elevated susceptibility to ligament injury, especially in the Caucasian population. Rs13946 polymorphism appears to increase the risk to tendon and ligament injuries. Rs71746744 and rs3196378 polymorphisms have a tendency to confer an elevated risk to MSTIs. However, no relevance is found between rs11103544 polymorphism and MSTIs. Given limitations in this meta-analysis, it is encouraged to verify these findings with complementary larger and well-designed prospective studies.


MTHFR) gene rs1801133 and rs1801131 polymorphisms influence MTHFR enzyme activity and plasma homocysteine concentration. In addition, variations in MTHFR functions likely play roles in the etiology of EH. Thus far, a large number of studies investigating the associations between the MTHFR polymorphisms and EH have provided controversial or inconclusive results. To better assess the purported relationship, we performed a comprehensive analysis of 52 published studies. Objective and Methods. Eligible studies were identified by searching the PubMed, Wanfang, and China National Knowledge Infrastructure (CNKI) databases. Odds ratios (ORs) with 95% confidence intervals (CIs) were estimated to assess the potential association between the MTHFR rs1801133 polymorphism and EH.  Essential hypertension (EH) is a common and multifactorial disorder that is likely to be influenced by multiple genes. The methylenetetrahydrofolate reductase , with similar results evident within race subgroups . Similar associations were also found in subgroups defined by the source of controls and genotype methods. To our regret, based on the limited studies, no association was detected for rs1801131 polymorphism.  Overall, 10712 patients and 11916 controls were involved; we observed significantly increased association between the MTHFR rs1801133 null genotype may increase EH risk. Future studies with larger sample sizes are warranted to evaluate this association in more detail. Our study provides evidence that the  Essential hypertension (EH) has a high prevalence rate worldwide and is considered to derive from complex interactions between diverse genes and environmental conditions , 2. The Most people with hypertension do not present with typical symptoms, making the condition easy to ignore and reducing the opportunities for early medical intervention. Therefore, the identification of high-risk patients is particularly meaningful, since identifying such patients with high blood pressure as early as possible allows for timely monitoring of blood pressure, regular follow-ups, and options for improving unhealthy lifestyles. In known high-risk patients, rising blood pressure can be immediately detected and treatment can be provided timeously, thus avoiding complications and, ultimately, reducing the incidence of hypertension and improving quality of life , 6, 7.MTHFR gene is a C to T transition at nucleotide position 677 (C667T) in exon 4, which results in a change from alanine to valine at amino acid 222. This mutation may lead to a decrease in MTHFR activity and heat tolerance and thus metabolic damage of Hcy, which may moderately increase plasma Hcy levels [MTHFR two polymorphisms may be related to EH development and susceptibility. It is therefore worthwhile to demonstrate whether there is an association between this polymorphism and EH risk, as it may provide guidance for the prevention and diagnosis of EH in the clinic. Thus far, numerous studies have reported the association between the MTHFR two common polymorphisms and EH risk. We therefore performed a comprehensive analysis of 52 different case-control studies to derive a convincing conclusion regarding this apparent association [Studies have shown that a high plasma concentration of homocysteine (Hcy) may injure the vascular endothelium, which results in hypertension and a predisposition toward atherosclerosis. Elevated Hcy levels have been identified as an independent risk factor for hypertension –11. Methy levels , 13; morociation .MTHFR two polymorphisms and EH susceptibility. All of the included studies met the following criteria: (1) association between MTHFR two polymorphisms and EH risk; (2) case-control study; (3) each genotype frequency is shown in tables; and (4) genotype distributions of the control were consistent, with a Hardy-Weinberg equilibrium (HWE) more than 0.05. Otherwise, studies should be excluded with the following issues: (1) no control, (2) incomplete genotype frequency data, (3) duplication studies, and (4) not according with HWE in control groups.Searches were performed in the PubMed, Wanfang, and China National Knowledge Infrastructure (CNKI) databases  using the following related keywords: polymorphism/variant/mutation, hypertension/essential hypertension, and MTHFR/methylenetetrahydrofolate reductase. We included all studies that described a relationship between the We collected the following information in our analysis: first author's last name, year of publication, ethnic origin, sample size (cases/controls), study design, HWE of controls and genotype methods, and genotype frequencies in cases/controls.In this meta-analysis, the quality was assessed using the Newcastle-Ottawa Scale (NOS) for cross-sectional study quality assessment. The methodological quality of each study , comparability, and outcome were checked using the NOS tool. Studies with a score of more than 7 out of 10 were considered as good quality. This cut-off point was determined after reviewing relevant meta-analyses from the literature –17. MoreMTHFR rs1801133 and rs1801131, we investigated the relationship between genetic variants and EH risk in five different models .The extracted data were imported into the Stata software program  for analysis. Odds ratios (ORs) with 95% CIs were used to measure the strength of the association , 20. TheQ test (chi-square) and I2 (%) statistical analysis. A fixed effect model was applied when the effects were assumed to be homogenous ; otherwise, the random-effect model was adopted  [The evaluation of heterogeneity within the included studies was assessed using Cochrane's 2 ≥ 50%) , 22. WheMTHFR polymorphism frequencies from expectations under HWE were assessed in controls using the Pearson chi-square test [http://biostat.mc.vanderbilt.edu/wiki/Main/PowerSampleSize#Windows) [The presence of potential publication bias was determined using the Egger/Begg's test and presented graphically in the form of a funnel plot . In addiare test . AnotherWindows) .MTHFR gene was conducted using the following analyses: polymerase chain reaction-restriction fragment length polymorphism (PCR-RFLP), TaqMan, sequencing, PCR, amplification refractory mutation system-PCR (ARMS-PCR), and high-resolution melt (HRM) genotyping.Genotyping to identify single nucleotide polymorphisms (SNPs) in the Random-effect metaregression analysis was performed to define the source of publication bias, with the subgroups of publication year, ethnicity, source of control, and genotype method set as independent variables and the log values considered as dependent variables .MTHFR and its potential functional partners in EH, we used the STRING online server (http://string-db.org/) to construct a MTHFR gene-gene interaction network [In order to fully understand the role of  network , 28.P < 0.05). The distribution of genotypes in all the controls was in agreement with HWE. In addition, we confirmed the minor allele frequency (MAF) reported for the seven main worldwide populations in the 1000 Genomes Browser [https://www.ncbi.nlm.nih.gov/snp/rs1801133), as follows: Global (0.335), European (0.345), East Asian (0.328), South Asian (0.167), African (0.123), African American (0.125), and Asian (0.265)  , we identified 364 articles from PubMed, 56 from CNKI, and 362 from the Wanfang database. In total, 552 articles were excluded after review of the abstract, leaving 230 articles for full article evaluation. Among them, 39 articles featured systematic analysis/meta-analysis/review; 24 covered only case groups; 17 articles were duplicates of other papers; 35 had no original numbers for case/control groups and presented only total numbers; 28 articles focused on H-type hypertension; 4 were related to aortic hypertension; and 27 covered hypertension in pregnancy . After e Browser  (https:/ (0.265) . When we (0.265) . Finally (0.265) . This pog/home/) . We alsog/home/) .MTHFR, based on 10,712 EH cases and 11,916 matched controls. We observed an increased association between the MTHFR rs1801133 polymorphism and EH in total population groups . In order to analyze the source of controls and determine the source of heterogeneity, the ratios were calculated for the HB and PB subgroups. The results showed significantly increased relationships in these groups  , PCR-RFLP  . As diffBegg's funnel plot and Egger's test were performed to determine the publication bias of the included studies. Significant obvious evidence of publication bias was detected in five genetic model analyses  with respect to ethnicity, source of control, and genotype methods, with a regression coefficient of 0.001, 0.004, 0.010, and 0.002, respectively. There was no association with publication year, which suggests that the heterogeneity from the rs1801133 polymorphism in EH may be due to the ethnicity, source of control, and genotype method subgroups Figures .MTHFR interacts with numerous genes. The ten most significant genes from the network of gene-gene interactions have been listed in Figures MTR), thymidylate synthase (TYMS), C-1-tetrahydrofolate synthase (MTHFD1), serine hydroxymethyltransferase 1 (SHMT1), serine hydroxymethyltransferase 2 (SHMT2), bifunctional methylenetetrahydrofolate dehydrogenase (MTHFD2), probable bifunctional methylenetetrahydrofolate dehydrogenase (MTHFD2L), aminomethyltransferase (AMT), and methionine synthase reductase (MTRR).Our analysis using the STRING online server indicated that While the precise causes of hypertension are still unknown, the risk factors include genetic factors, age, and unhealthy lifestyle practices, with 70-80% of hypertensive cases resulting from unhealthy lifestyle practices. As the risk factors for high blood pressure accumulate, the risk of high blood pressure increases , 31.The detection of significant polymorphisms may be a suitable method to predict the risk of hypertension in susceptible individuals. Our current study focused on EH, a common type of hypertension, and included 10,712 patients with EH and 11,916 healthy individuals. In the overall analysis, we observed that individuals carrying TT or a T-allele may have an increased risk of developing EH compared to those with CC or a C-allele (between 37% and 89%). In other words, individuals carrying T-allele or TT genotype may have more possibility to suffer from hypertension in a current or in future time. This can give these people warnings, such as regular changes in blood pressure, changing bad habits, moderating physical exercise, or early medical intervention. Significant heterogeneity was indicated in all of the genetic models. To determine the source of the variation, we analyzed the associations in other subgroups, such as ethnicity, source of control, and genotype method. In parallel, significant relationships were also observed with the ethnicity, source of control, and genotype method subgroups, which provided further support for hypothesis that the rs1801133 polymorphism is a risk factor for EH. In addition, when we used metaregression analysis to evaluate the source of heterogeneity, the aforementioned three subgroups emerged as significant sources of variation. The power of our study was 1, suggesting that our conclusions were accurate.MTHFR and related genes to elucidate potential functional interactions. Despite these disadvantages, it should be noted that the conclusions from our current study are similar to those from previous meta-analyses.Several meta-analyses regarding the rs1801133 polymorphism and hypertension have been published to date. For example, Wu et al. included 30 case-control studies, and their findings supported a role for the rs1801133 polymorphism in the risk of developing EH . Kosmas MTHFR. The average scores of the ten most significant genes were more than 0.9. The top three genes that were suggested to interact with MTHFR were MTR (0.995), TYMS (0.992), and MTHFD1 (0.989). MTHFR and MTR both participate in homocysteine metabolism, regulating different stages of the process. MTHFR converts 5,10-methylene-THF into 5-methyl-THF, while MTR catalyzes the demethylation of 5-methyl-THF to THF and the remethylation of homocysteine to methionine [A key feature of our study was the identification of gene-gene interactions for thionine , 37. Thethionine .Some limitations in our meta-analysis should be considered. First, our analyses indicated that the heterogeneity identified in our study emerged from ethnicity, source of control, and genotype methods. Future studies should optimize the design of both retrospective and prospective research projects to overcome this deficiency. Second, EH is a complex disease including genetic and other factors  , and futMTHFR 677T null genotype is associated with increased risk of EH, suggesting that further, well-designed large studies are necessary to confirm this relationship. Furthermore, it will be important to focus on the mechanism of action of the rs1801133T-allele to explain the complete chain of evidence for the prevention of EH in the future.Our meta-analysis provides evidence that the


The primary objective of this systemic review and meta-analysis was to investigate the risk of developing composite outcome of all cancers, regardless of the type of cancer among men with infertility diagnosis compared to fertile counterparts. The secondary objective was to compare the pooled risk of developing individual specific cancers between two groups.A systematic literature search was performed on the databases of PubMed (including Medline), Scopus, and Web of Science to retrieve observational studies published in English language from 01.01.1990 to 28. 02. 2021. They assessed cancer events in males with an infertility diagnosis compared to controls without infertility. The outcomes of interest were a composite outcome of cancers including all known cancer types, and also specific individual cancers. The fixed/random effects model was used to analyze heterogeneous and non-heterogeneous results. Publication bias was assessed using the Harbord test, Egger test, Begg test, and funnel plot. The pooled odds ratio of cancers was calculated using the DerSimonian and Laird, and inverse variance methods. Studies’ quality and risk of bias were assessed using structured standard tools.We included eight cohort studies involving 168,327 men with the diagnosis of infertility and 2,252,806 men without it. The total number of composite outcome of cancers as well as individual cancers including prostate, testicular and melanoma were 1551, 324, 183 and 121 in the infertile men and 12164, 3875, 849, and 450 in the fertile men, respectively. The pooled OR of the composite outcome of cancers, regardless of the type of cancer, in men with infertility was 1.4 folds higher than those without infertility . Meta-analysis of individual cancers including prostate, testicular and melanoma between two groups was carried out. The pooled ORs of testicular and prostate cancers in men with the diagnosis of infertility were significantly higher than controls without infertility . Additionally, the pooled OR of melanoma in men with infertility was 1.3 folds higher than those without infertility .A greater risk of cancers in men with male infertility was found suggesting that the history of male infertility might be an important risk factor for developing cancers in later life. Further well-designed long-term population-based prospective studies, considering all known cancers and their accompanying risk factors should be conducted to support our findings. Infertility is defined as failure to achieve a clinical pregnancy after 12 months or more of regular unprotected sexual intercourse . Althougper se, may play as a risk factor for the development of genitourinary cancers in men  , to achito study the pooled risk of developing composite outcome of all cancers, regardless of the type of cancer among men with the diagnosis of infertility compared to fertile men;to compare the pooled risk of developing individual specific cancers between men with the diagnosis of infertility and fertile men.The review question was framed using the PICO  statement as follows: P: men with the diagnosis of male infertility; I: risk of developing cancer; C: men without male infertility; O: overall and individual cancer. Study protocol was developed before the study and was used as the guideline to conduct this research , Scopus, and Web of Science to retrieve observational studies published in English language from 01.01.1990 to 28.02.2021. They should have investigated the risk of the development of cancer in males with the diagnosis of infertility. Further, a manual search in the references list of selected studies and other relevant reviews was carried out to maximize the identification of eligible studies. The following keywords, alone or in combination, were used during the search process:  AND  ; definition of infertility; follow-up period; quality assessment; outcome measurements in terms of the number and prevalence of cancer. Any disagreement in the selection of studies was resolved through holding discussions between the authors and also seeking comments from the third review author. The data extraction process was double-checked to ensure the accuracy of data collection before the meta-analysis and prevent bias in data extraction and data entry.Primary outcome of interest was the pooled risk of developing composite outcome of all cancers, regardless of the type of cancer among men with the diagnosis of infertility. Secondary outcome was the pooled risk of developing individual specific cancers among men with the diagnosis of infertility. Male infertility was defined as men reporting the experience of infertility with the duration of more than one year .Quality of the included studies was critically appraised in terms of the methodological structure and presentation of results. Two authors were made blind to the study’s author, country, and the journal’s title to evaluate the quality of each study independently. The quality of observational studies was assessed using the modification of the Newcastle–Ottawa Quality Assessment Scale for Non-Randomized Studies (NRS) . This scThe Risk Of Bias in Non-randomized Studies (ROBINS) tool in non-randomized studies of interventions and observational studies was used to assess risk of bias  as the C2) statistics and values above 50% were interpreted as heterogeneity. Given the heterogeneous results of the included studies, the pooled effect was calculated using the random effect model. Publication bias was assessed through the visual inspection of funnel plot, Harbord test, Egger test, and Begg test. When the funnel plot is symmetrical and the p values of Harbor test, Begg test, and Egger test are >0.05, no significant publication bias exists in the meta-analysis. Pooled OR (95% CI) was estimated using the DerSimonian and Laird, and inverse variance methods. Forest plots were drawn to show the estimation of pooled OR (95% CI) in the included studies. Sensitivity analysis was run to investigate the influence of each individual study on the estimation of overall meta-analysis summary. The graph of the results of an influence analysis in which the meta-analysis was re-estimated, omitting each study in turn, was drawn. Significant level was considered p <0.05 and all statistical analyses were performed using the STATA software .Meta-analysis was performed to evaluate the pooled OR (95% CI) of the outcomes of interest including individual cancers and composite of all cancers. Heterogeneity was evaluated using the I-squared    , whereas heterogeneity was observed among the studies with regard to the analysis of the composite outcome of all cancers and prostate cancer   Table 2.According to the results of publication bias tests, no substantial publication bias for meta-analysis was observed (Table 2), which also was confirmed by the symmetric funnel plot  . HoweverOur study have some limitations that should be considered in the interpretation of data. In our review, a considerable heterogeneity between the studies that assessed the risk of prostate cancer was observed. Different criteria for the screening and recruitment of infertile men in the included studies, lack of uniform data regarding risk stratification such as the grade and stage of cancers, and absence of risk factors such as race/ethnicity, socio-economic and demographics may have led to the heterogeneity. Therefore, definitive conclusions should be taken cautiously. Also, the included studies had large sample sizes in case and control groups, but the number of studies assessing the occurrence of cancer in men with infertility was limited and hindered us to run various subgroup analyses for the different types of cancers, stages of cancer, and the degree of male infertility impairment. Moreover, various confounders for the development of cancer such as lifestyle factors  could noThis systematic review and meta-analysis demonstrated an increased risk of composite outcome of cancers as well as melanoma, testicular and prostate cancers in men with the diagnosis of male infertility. Well-designed long-term prospective studies, considering all known cancers and their accompanying risk factors should be conducted to support our findings.The original contributions presented in the study are included in the article/SB-G conceptualized the study and was involved in study design, search on databases, study selection, data extraction, drafting the manuscript, and revising it critically for important intellectual content. MV was involved in study design, study selection, data extraction, manuscript drafting, editing, and revising it critically for important intellectual content. RB-Y contributed to quality assessment, data analysis, interpreting data and drafting the manuscript. MP contributed to quality assessment, data analysis, and interpreting data. All authors contributed to the article and approved the submitted version.Nord University, Bodø, Norway covered the article processing charge.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


Lin et al. recently did a network meta-analysis based on cardiovascular (CV) outcome trials (CVOTs) of sodium-glucose cotransporter 2 inhibitors (SGLT2is) and those of glucagon-like peptide-1 receptor agonists (GLP1RAs). Due to the absence of CVOTs directly comparing SGLT2is with GLP1RAs, Lin et al.'s network meta-analysis identified the indirect evidence that SGLT2is vs. GLP1RAs reduced hospitalization for heart failure (HHF) but did not reduce CV death and all-cause mortality (ACM) in patients with type 2 diabetes (T2D). We did another meta-analysis incorporating those CV outcome cohort studies directly comparing SGLT2is with GLP1RAs, and identified that SGLT2is vs. GLP1RAs were significantly associated with the lower risks of not only HHF but also CV death and ACM. These findings may suggest that SGLT2is should be considered over GLP1RAs in terms of preventing CV and all-cause death and HHF in T2D patients. Diabetologia. By performing network meta-analysis based on 21 placebo-controlled cardiovascular (CV) outcome trials (CVOTs), Lin et al.  for cohort studies .This meta-analysis was conducted according to the preferred reporting items for systematic reviews and meta-analyses (PRISMA) statement . Its stu studies . Any agrIn this meta-analysis we included 9 large PSM cohort studies –12. EachThis meta-analysis is the first one that provided the direct evidence regarding the relative efficacy of SGLT2is vs. GLP1RAs on death and cardiorenal endpoints in T2D patients by incorporating large PSM cohort studies directly comparing SGLT2is with GLP1RAs. Consistent with the indirect evidence from Lin et al.'s network meta-analysis  was observed in the meta-analysis on HHF. Although we utilized the random-effects model to derive the conservative pooled results, it would be beneficial that future studies could perform relevant subgroup analyses for this outcome to explore the sources of heterogeneity.Although this meta-analysis provided the direct evidence regarding the relative cardiorenal efficacy of SGLT2is vs. GLP1RAs, the evidence derived from cohort studies, which involve more risks of biases than randomized trials do. Although the cohort studies included in this meta-analysis performed PSM analysis to adjusted lots of confounding factors, there were probably some omissive factors. Thus, there is a need for CVOTs comparing SGLT2is with GLP1RAs in T2D patients, to further confirm the direct evidence revealed in this meta-analysis. Although no substantial heterogeneity was observed in the meta-analyses on most of the endpoints assessed in this study, the substantial heterogeneity (Lin et al.  revealedThe original contributions presented in the study are included in the article/MQ: design and writing manuscript. MQ, X-BW, and WW: conduct, data collection, and analysis. All authors contributed to the article and approved the submitted version.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


Bladder cancer (BCa) is a common cancer in North America and Europe that carries considerable morbidity and mortality. A reliable biomarker for early detection of the bladder is crucial for improving the prognosis of BCA. In this meta-analysis, we examine the diagnostic role of the angiogenin (ANG) protein in patients' urine with bladder neoplasm. th October 2020 databases. Meta-Disc V.1.4 and Comprehensive Meta-Analysis V.2.2 software calculated the pooled specificity, sensitivity, area under the curve (AUC), diagnostic odds ratio (DOR), positive likelihood ratio (LR+), negative likelihood ratio (LR−), Q∗ index, and summary receiver-operating characteristic (SROC) for the role of ANG as a urinary biomarker for BCa patients.  We performed a systematic literature search using ScienceDirect, Web of Science, PubMed/MEDLINE, Scopus, Google Scholar, and Embase, up to 10+ of 3.34 (95% CI: 2.02–5.53), LR− of 0.37 (95% CI: 0.32–0.44), DOR of 9.99 (95% CI: 4.69–21.28), and AUC of 0.789 and Q∗ index of 0.726 demonstrate acceptable diagnostic precision of ANG in identifying BCa.  Four case-control studies were included with 656 participants (417 cases and 239 controls) in this meta-analysis. The pooled sensitivity of 0.71 (95% CI: 0.66–0.75), specificity of 0.78 (95% CI: 0.73–0.81), LR This meta-analysis showed that ANG could be a fair biomarker for the diagnosis of BCa patients. The prevalence of BCa is increasing around the world, especially in developed countries ountries , alcoholountries , aromatiountries , nitrateountries , and heaountries , 15. BCaountries .Noninvasive tools for diagnosis or prediction of BCa have been broadly examined in recent years. Today, urine tests and urinary cytology (UC) tests are used as the primary diagnostic markers of bladder cancer. Biochemical and molecular investigations of blood and urine make a liquid biopsy that could suggest new approaches for prevention, monitoring, and diagnosis . Biomarkα-helices, seven β-sheets, and three disulfide bonds. The gene encoding ANG is located on chromosome 14q11.2  The level of ANG protein in the urine was measured. [4] The diagnostic value or prognostic significance of ANG in BCa patients was assessed. [5] The true-positive, true-negative, false-positive, and false-negative values were reported or obtained by calculating ROC curve data.The studies were excluded when they were (1) studies without available data or incomplete information, [2] duplicate publications, [3] not written in English, [4] case reports, [5] meta-analyses, [6] letters to editors, [7] comments, [8] reviews, [9] unrelated studies that were eliminated through careful review of the title and abstract of each paper, and [10] articles with only the abstract of the article available.AH. A, H. A, and M. M independent reviewers screened the full texts and judged their quality. Conflicts were resolved by discussion to ensure compatibility. The following data was captured in a predesigned form: first author, year of publication, type of sample, sample size, country, method of detection, and results.The Quality Assessment of Diagnostic Accuracy Studies QUADAS-2) checklist assessed each study's quality in this paper with nine questions , negative likelihood ratio (LR−), and diagnostic OR (DOR) with 95% confidence interval (CI), based on the extracted data of true-positive, false-positive, true-negative, and false-negative.The diagnostic value of the ANG protein evaluated in patients with BCa was assessed with the random effects model by MetaDisc V.1.4  and fixed effects model by Comprehensive Meta-Analysis V.2.2  software to summarize the pooled specificity, sensitivity, positive likelihood ratio (LRQ and I-squared (I2) estimated heterogeneity between studies. The area under the “summary receiver-operating characteristic” (SROC) curves described the overall performance detection method.Spearman's rank correlation coefficient is a method that can be used to summarize the strength and direction (negative or positive) of an associated with two variables and also was used to evaluate cut-off threshold effects between sensitivity and specificity. The Cochran's +, LR−, and sample size are listed in p value of 0.40 (p > 0.05) suggested no heterogeneity from the threshold effect. However, it showed a moderate positive correlation.Results have shown for metaregression and Spearman rank correlation of sensitivity (1 – specificity) to assess the threshold effect in all test accuracy studies included in the meta-analysis . The ranI2 heterogeneity of sensitivity, specificity, LR+, LR−, and DOR were 20.8%, 92.1%, 82.7%, 0%, and 77.4%, respectively  for sensitivity, 37.95 (p < 0.0001) for specificity, 17.33 (p = 0.001) of positive likelihood ratio, 53.67 (p = 0.521) of negative likelihood ratio, and 13.26 (p = 0.004) of diagnostic odds ratio, respectively  and 0.376 (95% CI: 0.320-0.441), respectively  and 0.780 (95% CI: 0.736-0.819), respectively Figures . The poo Figures . The poo Figures . The AUC Figures . The ROCIn recent years, numerous biomarkers have been reported, particularly those that play a role in cancer development and progression to diagnose BCa . A colleMicroRNAs (miRs) are small 18-25 nucleotide long nonprotein-coding RNAs that regulate gene expression by pairing to the 3′ untranslated region of their target mRNAs in body fluids as free circulating miRs –56. Urinin vivo and is associated with multiple human neoplasms, diabetic retinopathy, and arthritis. ANG performs a fundamental function in ribosome biogenesis and many cellular processes [ANG is an effective inducer of angiogenesis rocesses . Howeverrocesses . Besidesrocesses –64. This+ of 3.344 (2.021-5.533) demonstrated that the diagnostic accuracy of BCa was improved by 3.344-fold with the positive ANG results and the pooled LR-0.376 (0.320-0.441).The pooled LR+ and LR−, a significant diagnostic index called DOR can be calculated. The higher range of DOR value shows a better diagnostic performance from 0 to infinity limit [Q∗ index is set as the point of indifference on the ROC curve, where the sensitivity and specificity are equal [Q∗ index was found to be 0.726, and SROC was 0.789 area under the curve (AUC) for ANG in diagnosing BCa patients. These findings highlight the role of ANG in detecting BCa and potentially could have a crucial role in identifying patients with BCa.By merging the LRty limit . In thisre equal . In thisp = 0.004), acute myelogenous leukemia (AML) (p = 0.001), multiple myeloma (MM) (p < 0.001), myelodysplastic syndromes (MDSs) (p = 0.001), and heart failure (p < 0.001) than those in healthy individuals. However, patients with hepatocellular cancer (HCC) (p = 0 249), breast cancer (p = 0.443), non-Hodgkin lymphomas (NHLs) (p = 0.257), and melanoma (p = 0.550) did not have significantly higher serum ANG levels than healthy controls. They did not consider bladder cancers in their study. Overall, their study's findings show that serum ANG levels in healthy individuals usually remain in a specific range and are associated with various diseases. Serum ANG levels are not currently a clinical diagnostic marker for the disease. However, significant changes in serum ANG levels in cancers and CVDs suggest that this protein might be involved in the pathogenesis and could be a moderate biomarker for these diseases. Various malignant carcinomas, including prostatic cancer, depend on angiogenesis for growth, invasion, and progression. Pina et al. [p = 0.008). They concluded that serum angiogenin levels might help differentiate between cancer and noncancer patients among prostate biopsy candidates.In a meta-analysis reported by Yu et al. , the rola et al.  evaluateThis meta-analysis had some limitations. The major limitation was the small number of studies and participants included because most of the studies were designed for evaluating the diagnosis performance of other biomarkers such as microRNAs. On the other hand, data on the prediction of ANG was also often incompletely reported. Additionally, many trials were of small sample size, with only a few patients assessed by ANG. Hence, the results of this meta-analysis should be confirmed in a study with larger sample size. However, we showed the significant Spearman rank correlation between ANG and BCa patients. Although we demonstrated the diagnostic role of ANG in BCa patients, heterogeneity was low but still present in this analysis. Confounding factors such as gender, race, age, BMI, smoking, comorbidities, and pharmacotherapy may be other causes of heterogeneity that could have affected the results.This is the first meta-analysis to appraise the urinary ANG marker in patients with BCa and may have clinical value for screening bladder cancer to the best of the authors' knowledge.Eventually, this meta-analysis showed urinary ANG as a fair and noninvasive tumor marker that can detect BCa patients. Nonetheless, more extensive studies with a larger sample size are needed.


The presence of central lymph node metastases (CLNM) has been suggested as a risk factor for poorer prognosis and recurrence in papillary thyroid microcarcinoma (PTMC). However, the clinicopathologic factors for CLNM in clinical node-negative (CN0) PTMC were not well defined. This study aimed to perform a systematic review and meta-analysis to investigate the significant clinicopathologic predictors of CLNM in CN0 PTMC.A systematic literature search was performed in PubMed, Embase, Cochrane Library, and Web of Science. Case-control studies on the association of clinicopathologic risk factors with CLNM in CN0 PTMC were included.CI: 1.49–2.87, P < 0.001), multifocality , tumor size > 5 mm , and extrathyroidal extension  are significantly associated with increased risk of CLNM in CN0 PTMC. A sample size with a cutoff point of 200 was identified as the source of heterogeneity for sex according to meta-regression . Then, the subgroup analysis of male was performed, which illustrated that male increased the risk of CLNM in the small sample group (SG) and the large sample group (LG) by 6.11-folds and 2.01-folds, respectively .Thirteen eligible studies involving 6068 patients with CN0 PTMC were included. From the pooled analyses, male The online version contains supplementary material available at 10.1186/s12957-022-02573-7. Papillary thyroid carcinoma (PTC) has received much attention over the past decade due to its rapid global increasing incidence, with an incidence of 4.3/100000 to 143.3/100000 in women in various populations worldwide . Owing tCRD42021242211).This systematic review an meta-analysis were conducted according to the guidelines proposed by the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) . Online A literature search was conducted from PubMed, Embase, Cochrane Library, and Web of Science for articles published until February 11, 2022. Two independent authors (XZ Wen and QM Jin) identified articles with a combination of the following terms: “Papillary Thyroid Microcarcinoma,” “Lymphatic Metastasis,” “Risk factors,” and “Case-Control Studies” with language restriction “English.” All subject words and related random words were used for the search. The detailed search strategies were documented in Supplementary file All articles obtained from the search were screened according to the inclusion criteria: (1) original English articles; (2) CN0 PTMC patients who received total thyroidectomy or lobectomy and CLND as the primary surgical, and the CN0 was defined as no preoperative clinical evidence of lymph node involvement by ultrasonography; (3) study of the association between CLNM and relevant clinicopathologic and/or ultrasonic risk factors; (4) multivariate regression analysis was used to return at least one statistically significant risk factor expressed by odds ratio (OR) and 95% confidence interval (CI); and (5) case-control study. The selection process was screened by titles and abstracts initially to exclude the duplicates, conference abstracts, reviewers, meta-analysis, and non-English articles, and then the full text of the remaining studies was downloaded and assessed independently by two investigators (XZ Wen and QM Jin) based on the above inclusion criteria. A third investigator (XX Cen) was introduced to resolve the disagreements.The following data were extracted by two investigators (XZ Wen and QM Jin) independently in standardized forms: first author, the published year, country, sample size, number and percentage of males, mean age at diagnosis, number and percentage of CLNM patients, type of surgery, tumor size, multifocality, extrathyroidal extension, and bilateral and capsule involvement. Disagreements were resolved through consultation with the third investigator (XX Cen).Each study’s risk of bias was assessed based on the Newcastle-Ottawa scale (NOS), which consists of three columns: selection, comparability, and exposure. A score of more than 6 stars was considered to be of high quality. Two investigators (XZ Wen and QM Jin) performed quality assessment independently, and the disagreements were resolved by discussing it with the third investigator (XX Cen).Cochran’s chi-squared and Higgins’ I2 statistic were applied to evaluate the heterogeneity, with considerable heterogeneity defined as P < 0.10 or I2 > 50%. When moderate or severe heterogeneity existed, the sensitivity analysis used metainf meta-based influence analysis tool, and meta-regression was performed to explore the sources of considerable heterogeneity. The risk of publication bias was analyzed using Egger’s test and funnel plot. When the publication bias was presented, the trim and filling analysis was performed to calibrate the bias. All statistical analyses were processed using ReVman software 5.3  and Stata software 16.0 . All images were created with Adobe Photoshop CS6 13.0  for Windows.The pooled ORs and 95% CIs of the significant risk factors for CLNM were calculated. The fixed-effect model or the random-effect model was chosen according to the heterogeneity. As shown in Fig. The characteristics of included studies are shown in Table According to the NOS scale, the risks of bias of the included studies were evaluated. Among the included studies, 8 studies scored 7 stars and the other 5 studies scored 8 stars. All included studies had more than 6 stars, which were considered of high quality. The specific scores are shown in Table P = 0.096 < 0.1; I2 = 42.3% < 50%), a random-effects model was applied to the pooled analysis. As shown in Fig. OR, 2.43; 95% CI, 2.43–3.22.; P < 0.001).There were 8 studies that revealed sex was an independent risk of CLNM in CN0 PTMC. Due to the moderate heterogeneity , a fixed-effects model was applied in the pooled analysis. As shown in Fig. OR, 1.88; 95% CI, 1.54–2.29; P < 0.001).Multifocality was investigated as a significant risk of CLNM in CN0 PTMC in 6 studies. Due to the low heterogeneity , a fixed-effects model was applied. As shown in Fig. OR, 1.84; 95% CI, 1.55–2.18, P < 0.001).Tumor size was identified as a significant independent risk factor for CLNM in CN0 PTMC in 10 studies. However, only 5 studies that investigated tumor size > 5 mm as an independent risk factor were included in the final pooled analysis owing to the consistent definition. Due to the low heterogeneity . As shown in Fig. OR, 2.38; 95% CI, 1.65–3.44; P < 0.001).ETE was identified as a significant independent risk factor for CLNM in CN0 PTMC in 6 studies. Unfortunately, one study was excluded due to the definition of gross ETE, which was defined as macroscopic tumor extension outside the thyroid gland based on physical examination. A random-effects model was applied in the final pooled analysis due to the considerable moderate heterogeneity  and ETE (P = 0.006). We then applied the trim and filling analysis to calibrate the biases of sex and ETE by filling three and two studies respectively, which eventually conducted the calibrated values of ORs of male and ETE were 2.07 and 1.96 .The publication bias was evaluated by the funnel plot and the Egger’s test Fig. . Unfortut = 3.18, P = 0.033), suggesting that the sample size might be the true source of heterogeneity for sex. However, no significant variables were found in the meta-regression of ETE.Based on the sensitivity analyses, country, type of surgery, and sample size were identified as suspicious sources of heterogeneity for sex and ETE by carefully reviewing the recorded variables of the included studies. Then the meta-regression was performed. Fortunately, only the sample size with a cutoff point of 200 was a significant variable in sex (n = 3) and large sample group (LG) with a size > = 200 (n = 5). Fixed-effect models were both applied in 2 subgroups due to the absence of heterogeneities . Moreover, significant difference in heterogeneity was observed between subgroups (P = 0.002). As shown in Fig. OR, 6.11, 95% CI, 3.16–11.81, P < 0.001; LG: OR, 2.01, 95% CI, 1.65–2.46, P < 0.001). Furthermore, significant publication biases were not founded in both subgroups.According to the sample size, the 8 included studies that investigated sex were divided into 2 subgroups: small sample group (SG) with a size < 200 (In contrast to the better long-term survival prognosis of PTC, it has an unacceptable high incidence of CLNM, even in CN0 PTMC. The previous studies revealed the incidence of CLNM was about 18.3–50% in CN0 PTMC , 25. FurTo the best of our knowledge, this is the first meta-analysis to assess clinicopathologic risk factors for CLNM in CN0 PTMC. In this study, we found that male gender, multifocality, tumor size > 5 mm, and ETE were statistically significant clinicopathologic risk factors for CLNM in CN0 PTMC. Consistent with our finding, numerous studies identified the above clinicopathologic risk factors that were associated with lymph node metastases , 27. Qu,It is worth noting that many clinicopathologic factors were excluded from the pooled meta-analysis, which were significant risk factors in the individual included studies due to the insufficient amounts of eligible researches and the inconsistent definition of the same clinicopathologic factors. For example, age was investigated as a significant independent risk factor for CLNM in 5 included studies of our meta-analysis. However, the definitions of age contained 45 years and 50 years. According to existing studies, age was associated with aggressiveness of PTC, such as CLNM, LLNM, vascular invasion, and ETE , 33. NieThis meta-analysis has some limitations. First, the available studies included in the specific final meta-analysis were relatively insufficient due to the limited number of existing studies on risk factors of CLNM in CN0 PTMC, especially for the subgroup analysis of the ETE and tumor location. Second, heterogeneities and publication biases were found in some risk factors, which may affect the stability of the conclusion. Fortunately, the source of the heterogeneity was identified in male, and the subgroup analysis addressed the defect. Moreover, the results of the trim and filling analysis suggested that the general direction of the conclusion was unchanged. In addition, some potential significant predictors of CLNM in CN0 PTMC, such as age, were not included in the meta-analysis due to the inconsistent definition of split points. Nevertheless, the predictive values of male, multifocality, tumor size > 5 mm, and ETE for CLNM in CN0 PTMC were confirmed by our meta-analysis. Further updating of our meta-analysis with more eligible studies is still needed to identify more reliable predictors of CLNM in CN0 PTMC. These limitations remain to be explored in subsequent work.In summary, this systematic review and meta-analysis showed that male, multifocality, tumor size > 5 mm, and ETE were significantly associated with a higher risk of CLNM in CN0 PTMC. Specifically, compared to female, solitary, tumor size ≤ 5mm, and non-ETE PTMC, male, multifocality, tumor size > 5mm, and ETE increase 2.07-folds, 1.88-folds, 1.84-folds, and 1.96-folds risk of CLNM in CN0 PTMC, respectively. Male, multifocality, tumor size > 5 mm, and ETE may be reliable clinical predictors for CLNM in CN0 PTMC. Moreover, Prophylactic CLND should be considered in the surgical decision-making of CN0 PTMC patients, who are male, multifocal, with tumor size > 5 mm, and with ETE.Additional file 1: Table S1. PubMed search strategy. Table S2. Embase search strategy. Table S3. Cochrane Library search strategy. Table S4. Web of Science search strategy.


It is raised in various cardiovascular diseases, but its value in predicting disease severity or mortality outcomes has been controversial. Therefore, we conducted a systematic review and PubMed and Embase were searched until 23rd June 2021 for studies that evaluated the relationship between sST2 levels and cardiovascular disease severity or mortality.meta-analysis. In acute heart failure, sST2 levels did not differ between survivors and non-survivors . Elevated sST2 levels tend to be associated with increased mortality risk . In chronic heart failure, sST2 levels were higher in non-survivors than in survivors  and elevated levels were associated with increased mortality risk . sST2 levels were significantly higher in severe disease compared to less severe disease . Finally, in stable coronary artery disease, sST2 levels were higher in non-survivors than survivors  and elevated levels were significantly associated with increased mortality risk .A total of 707 entries were retrieved from both databases, of which 14 studies were included in the final sST2 significantly predicts disease severity and mortality in cardiovascular disease and is a good predictor of mortality in patients with stable coronary artery disease and chronic heart failure. PubMed and EMBASE were searched for studies that investigated the relationship between soluble suppression of tumorigenicity-2 (sST2) levels and cardiovascular diseases using the following terms: [(Soluble suppression of tumorigenicity-2 OR sST2) and ]. The databases were searched from inception to 23rd June 2021, with no language restrictions. The following inclusion criteria were applied: i) case-control, prospective or retrospective cohort studies in humans; ii) sST2 values were provided and related to disease severity or mortality in cardiovascular diseases; iii) the study assessed cardiovascular diseases. The following studies were excluded: i) did not assess cardiovascular diseases; ii) systematic reviews, meta-analyses and editorials; iii) absence of comparable data.This systematic review and Supplementary Table 1.Quality assessment was performed using the Newcastle–Ottawa Quality Assessment Scale (NOS). The point score system evaluated the categories of 1) study participant selection, 2) comparability of the results, and 3) quality of the outcomes. The following characteristics were assessed: a) representativeness of the exposed cohort; b) selection of the non-exposed cohort; c) ascertainment of exposure; d) demonstration that outcome of interest was not present at the start of study; e) comparability of cohorts on the basis of the design or analysis; f) assessment of outcomes; g) follow-up period sufficiently long for outcomes to occur; and h) adequacy of follow-up of cohorts. Each criteria met contributed to a point in the scale, which varied from zero to nine points: Study quality was deemed poor if<5 points, fair if 5 to 7 points, and optimal if 8 or more points. The details of the NOS quality assessment are shown in 2.2meta-analysis, the extracted data elements consisted of: surname of first author, publication year, study design, follow-up duration, sample size, gender, age, and cut-off point for sST2 levels. Two reviewers (CI and KSL) independently reviewed each included study and disagreements were resolved by adjudication with input from a third reviewer (GT).Data extraction was performed with a pre-specified spreadsheet in Microsoft Excel. All publications identified were assessed for compliance with the inclusion criteria. In this meta-analysis. For the relationship between sST2 and mortality, multivariate adjusted hazard ratios (HR) with 95% confidence interval (CI) were extracted and analyzed for each study. When values from multivariate analysis were not available, those from univariate analysis were used. When the latter were not provided, raw data where available were used to calculate unadjusted risk estimates.Mean differences in sST2 levels between survivors and non-survivors were extracted from each study and subsequently pooled in our 2I statistic from the standard chi-square test, which describes the percentage of the variability in the effect estimates resulting from heterogeneity. A value of 2I > 50% was considered to reflect significant statistical heterogeneity. The random-effects model using the inverse variance heterogeneity method was used if 2I > 50%. To locate sources of heterogeneity, sensitivity analysis was performed by excluding one study at a time. Subgroup analyses based on different disease conditions and different endpoints were performed. Funnel plots showing standard errors or precision against the logarithms of the odds ratio were constructed. The Begg and Mazumdar rank correlation test and Egger’s test were used to assess for possible publication bias.Heterogeneity across studies was determined using Cochran's Q value, which is the weighted sum of squared differences between individual study effects and the pooled effect across studies; and the 3meta-analysis The flow diagram detailing the search strategy and study selection process is shown in 3.1meta-analysis because it did not provide any measure of dispersion Fig. 2A). Pooled meta-analysis showed a mean difference of 24.2 ng/ml (standard error: 13.0 ng/ml) but this did not quite reach statistical significance (P = 0.06). I2 took a value of 95%, indicating the presence of substantial heterogeneity. The funnel plot of standard error against mean difference is shown in Supplementary Figure 1. Begg and Mazumdar rank correlation suggested no significant publication bias . Egger’s test demonstrated no significant asymmetry . Sensitivity analysis excluding one study at a time did not significantly affect the pooled estimate  from five studies assessing acute heart failure were analysed estimate .Fig. 2MeB). Elevated levels of sST2 increased the risk of all-cause mortality by 12% but this value was not statistically significant . I2 took a value of 88%, indicating substantial heterogeneity. The funnel plot is shown in P < 0.05). Sensitivity analysis excluding one study at a time did not significantly affect the pooled estimate . Nevertheless, pooled results showed a mean difference of 0.19 ng/ml . I2 took a value of 0%, indicating absence of heterogeneity. The funnel plot of standard error against mean difference is shown in P > 0.05). Sensitivity analysis excluding one study at a time did not significantly affect the pooled estimate  from six studies for chronic heart failure were analysed Fig. 2estimate .D). Our meta-analysis showed that elevated levels of sST2 significantly increased the risk of all-cause mortality by 64% . I2 took a value of 82%, indicating the presence of substantial heterogeneity. The funnel plot for standard error against mean difference is shown in P < 0.01). Sensitivity analysis excluding one study at a time did not significantly affect the pooled estimate . Nevertheless, pooled meta-analysis showed a mean difference of 1.56 ng/ml . I2 took a value of 98%, indicating the presence of substantial heterogeneity. The funnel plot of standard error against mean difference is shown in P > 0.05). Sensitivity analysis excluding one study at a time did not significantly affect the pooled estimate . Pooled meta-analysis showed a mean difference of 2.96 ng/ml . I2 took a value of 80%, indicating the presence of substantial heterogeneity. A funnel plot of standard error against mean difference is shown in P > 0.05). Sensitivity analysis excluding one study at a time did not significantly affect the pooled estimate  from three studies on stable coronary artery disease were included Fig. 3estimate .Fig. 3MeB). Our meta-analysis showed that elevated levels of sST2 significantly increased the risk of all-cause mortality by 32% . I2 took a value of 57%, indicating the presence of moderate heterogeneity. The funnel plot is shown in Supplementary Figures 13. Begg and Mazumdar rank correlation suggested no significant publication bias . Egger’s test demonstrated significant asymmetry . Sensitivity analysis excluding one study at a time did not significantly affect the pooled estimate  and stable coronary artery disease .The main findings of this systematic review and Novel biomarkers have emerged and played an important role in assessing and monitoring the risk of patients with cardiovascular events in recent years. For example, natriuretic peptides (NPs) are produced in response to myocardial stress; mid-regional pro-adrenomedullin (MR-proADM) is related to global stress; and ST2 reflects ventricular fibrosis and remodeling Circulating sST2 levels are increased in response to inflammatory diseases and heart diseases The cardioprotective IL-33/ST2L signaling pathway prevents the myocardium from maladaptive hypertrophy, fibrosis and cardiomyocytes apoptosis, reducing cardiac dysfunction and improving survival Inflammation has been recognized to play an important role in the pathogenesis of different cardiovascular diseases. Higher concentrations of sST2 were associated with disease progression and predicted prognosis. As its concentration is less prone to haemodynamic fluctuations, it may mirror the progress of myocardial remodeling 5meta-analysis was to examine possible associations between sST2 levels and mortality in different cardiovascular conditions, it did not aim to provide risk quantification. Moreover, it did not examine sensitivities and specificities of sST2, which would be important for assessing its diagnostic ability. Future studies can perform meta-analysis of these statistical measures to explore this further. Therefore, the value of sST2 should be interpreted in the clinical context of the patient in conjunction with other laboratory test results for the purposes of risk stratification and deciding appropriate management.This study has many strengths. Firstly, all included studies had a prospective design, so our results are more accurate and less prone to biases. Secondly, most of the included studies used multivariate analysis, reducing the influence of confounders. However, several limitations should be acknowledged. There was a high degree of heterogeneity observed. This may be due to differences in quantification assays used for sST2 and country of origin. Despite the promising predictive value of sST2 in predicting mortality, we did not define a specific cut-off value due to the heterogeneity among studies and selection criteria. Whilst the goal of this 6High sST2 levels predict mortality in chronic heart failure and in stable coronary heart disease. It therefore provides incremental value for risk stratification purposes in these patient populations.The authors declare that they have no known competing financial interests or personal relationships that could have appeared to influence the work reported in this paper.


Evidence associating diet with the incidence of renal cell carcinoma (RCC) is inconclusive. We aimed to summarize evidence associating dietary factors with RCC incidence and assess the strength and validity of this evidence.We conducted an umbrella review of systematic reviews or meta-analyses (SRoMAs) that assessed the association between diet and RCC incidence. Through April 2021, PubMed, Web of Science, Embase, The Cochrane Library, Scopus, and WCRF were searched. Two independent reviewers selected studies, extracted data, and appraised the quality of SRoMAs. According to credibility assessment criteria, evidence can be divided into five categories: convincing (class I), highly suggestive (class II), suggestive (class III), weak (class IV), and nonsignificant (class V).n = 6), foods (n = 13), beverages (n = 4), alcohol (n = 7), macronutrients (n =15), and micronutrients (n =19). No meta-analyses had high methodological quality. Five meta-analyses exhibited small study effects; one meta-analysis showed evidence of excess significance bias. No dietary factors showed convincing or highly suggestive evidence of association with RCC in the overall analysis. Two protective factors had suggestive evidence  and vitamin C ) in overall analysis. One protective factor had convincing evidence ) in Europe and North America and one protective factor had highly suggestive evidence ) in North America.Twenty-nine meta-analyses were obtained after screening. After excluding 7 overlapping meta-analyses, 22 meta-analyses including 502 individual studies and 64 summary hazard ratios for RCC incidence were included: dietary patterns or dietary quality indices  in the overall analysis. Increased intake of vegetables and vitamin C is negatively associated with RCC risk. Moderate drinking might be beneficial for Europeans and North Americans, and cruciferous vegetables might be beneficial to North Americans, but the results should be interpreted with caution. More researches are needed in the future.CRD42021246619PROSPERO The online version contains supplementary material available at 10.1186/s12916-021-02229-5. Renal cell carcinoma (RCC) is the most fatal cancer of the urinary system and accounts for approximately 3% of all adult malignancies , 2. The Treatment of RCC results in a heavy financial burden. According to the 2017 French health insurance database, the cost of targeted therapy for patients with advanced RCC is €5546/person/month . TherefoNumerous studies have shown that diet is inextricably linked to the tumorigenesis of RCC –14. EvidTo summarize and assess the strength and validity of the available evidence and to evaluate its quality, an umbrella review of SRoMAs of observational studies was conducted to investigate the associations between various dietary factors  and the incidence of RCC. A systematic evaluation of the available data from published SRoMAs on a particular research issue is referred to as an umbrella review –32. UmbrThe protocol was registered in PROSPERO (registration number: CRD42021246619), and our search strategy followed the PRISMA 2020 statement: an updated guideline for reporting systematic reviews .The inclusion criteria were as follows: 1) dose-response or high vs low SRoMAs of observational studies ; (2) SRoMAs reporting odds ratios, relative rates, risk ratios, or hazard ratios as effect sizes, and SRoMAs including summary hazard ratios (SHRs) and 95% CIs. PICO definitions: (1) population: people of any sex, age, source of cases, country, and ethnicity; (2) intervention: dietary factors including various foods, dietary patterns, dietary quality index, beverages, alcohol, macronutrients, and micronutrients  individual study, (2) animal or experimental study, (3) conference literature, (4) editorial, (5) systematic review without meta-analysis , (6) meta-analysis of individual patient data, (7) exposures were plasma levels or biomarkers, and (8) diseases were other urologic tumors.PubMed, Web of Science, Embase, Scopus, and The Cochrane Library were systematically searched using a predefined search strategy from inception until April 2021. We also manually searched World Cancer Research Fund (WCRF) for their SRoMAs, which was rigorously conducted by their groups of experts. The language was limited to English. The literature search was conducted independently by two reviewers . Disagreements during the search process were resolved by consensus through discussion. A detailed search strategy for the Embase database has been provided in Additional file Study selection and data extraction were conducted by one author (ZCL) and double-checked by another author (ZTF). When a disagreement emerged, a consensus was reached through discussion. For each published SRoMA, data extracted included the title, first author, year of publication, exposures (including doses), number and types of individual studies included, type of effect model, sample size (participants and cases), type of meta-analysis metric, summary estimated effect  and 95% CI, type of comparison, quality appraisal tool for the individual study, publication bias, heterogeneity, funding information, and conflict of interest. For SRoMAs examining the associations between a specific dietary factor and multiple diseases or health outcomes, only data related to RCC were extracted.The following data were extracted from individual studies in a SRoMA: first author, publication year, exposures (including doses), sample size (participants and cases), summary estimated effect that adjusted for the most confounders and 95% CI, and adjusted covariates. According to the literature, the most important potential confounders in the investigation between dietary factors and the risk of RCC include age, sex, smoking, BMI, total energy intake, alcohol intake, and physical activity. We utilized the form of the Joanna Briggs Institute as our data extraction guide  utilizing the AMSTAR 2 online quality appraisal instrument and providing an overall rating for the SRoMAs included , 37. VerAMSTAR 2 is a reliable and validated methodological quality appraisal instrument with 16 items . Seven oSRoMAs are continually updated by incorporating relevant evidence from new studies to maintain validity and applicability. Otherwise, there would be a risk of inaccurate evidence or even a situation where erroneous evidence misleads decision-makers or researchers –42. MoreAs in the previous umbrella review , in the If the overlapping SRoMAs were all outdated, discarding them directly would result in inadequate data utilization, which would lead to a risk of bias and inaccurate results, so the outdated overlapping SRoMAs were selected according to the 3 principles mentioned above for managing contemporary overlapping SRoMAs.p<10-6 and p<0.05, respectively), the number of cases exceeded 1000 or the number of participants for consecutive outcomes exceeded 20,000, there was low heterogeneity (I2<50%), the 95% prediction interval (PI) did not include a null value, there was no excess significance bias, and there were no small study effects . Evidence was judged to be highly suggestive (class II) if the following criteria were met: statistically significant results were observed for both the association and the largest study ; and the number of cases exceeded 1000 or the number of participants for consecutive outcomes exceeded 20,000. Evidence was judged to be suggestive (class III) if the following criteria were met: associations with statistical significance were observed for the association (p<10-3); the number of cases exceeded 1000 or the number of participants for consecutive outcomes exceeded 20,000. Evidence was judged to be weak (class IV) if it met the criterion of a statistically significant association (p<0.05). The evidence was judged to be nonsignificant (class V) if no statistically significant association was observed (p≥0.05)As in previous umbrella reviews , 29, 30,We recalculated the SHR and corresponding 95% CI for the included SRoMAs for each exposure using the Sidik and Jonkman random-effects model , 51, whiIf both dose-response and high vs low SRoMAs were available, we recalculated the SHR and corresponding 95% CI only for the SRoMA with the highest number of included individual studies. If data related to the dose-response SRoMAs were missing, we calculated the SHR and corresponding 95% CI for the high vs low SRoMAs. If data could not be extracted from the SRoMAs, we could not recalculate the SHR and corresponding the 95% CI, and therefore, we directly used the SHR and 95% CI from the included SRoMAs.p value for each exposure [We considered that dietary factors were not associated with the incidence of RCC if the 95% CI included a null value or an SHR of 1. We used the meta package in R to calculate the SHR, 95% CI, and exposure .I2 to assess the heterogeneity of each published SRoMA. I2 values range from 0 to 100% and are used as a measure of the magnitude of heterogeneity among individual studies and can be used to present the percentage of the total variance caused by heterogeneity [I2 <50% can be considered nonsignificant heterogeneity. τ2 was also calculated, which can explain the between-study variance associated with risk estimates because I2 is affected by study size [I2, τ2, and 95% PI for each published SRoMA were calculated with the meta package in R [We used ogeneity . I2 <50%udy size . In addiudy size . The I2,age in R .For published SRoMAs, contour-enhanced funnel plots were used to assess whether indications of publication bias existed . The prep value<0.10 with a more conservative effect in the largest study  than in the random-effects meta-analysis was deemed to be indicative of small study effects. To estimate true effects, the “trim-and-fill” approach was applied when Egger’s test was statistically significant [Egger’s regression asymmetry test was employed to see if indications of small study effects existed  . As in pnificant . The connificant , 59.t distribution [p value<0.10 (two-tailed) for the observed number (O) > expected number (E) implies an excess significance bias, as described in previously published umbrella reviews [Based on a chi-square test, the excess significance test was used to assess whether the observed number of statistically significant studies (O) exceeded the expected number of statistically significant studies (E) , 61. Forribution . The exp reviews , 24, 27. reviews .Where sample sizes permitted, subgroup analyses were performed by region and study design. For associations with evidence levels of classes I to III, we performed sensitivity analyses by removing small-sized studies and low-quality studies which were defined in the included SRoMA. We conducted all statistical analyses in R (version 4.0.5).There were no patients involved in any part of this umbrella review, including establishing the research questions, outcome measures, study design, or implementation; interpretation of results; and suggestions for manuscript writing. We will disseminate the results through social media and relevant conferences.n=6), foods (n=13), beverages (n=4), alcohol (n=7), macronutrients (n=15), and micronutrients (n=19).A total of 1587 publications were identified in six databases, and 1029 publications were obtained after excluding duplicates. After title and abstract screening, 86 publications required full-text screening. Twenty-nine published meta-analyses were obtained after full-text screening , 64–88. We identified the following exposures associated with the incidence of RCC: healthy dietary pattern , unhealtn=298), 39% were case-control studies (n=196), and 2% were pooled studies (n=8).The overall characteristics of the 22 meta-analyses that were included in the umbrella review are summarized in Table n=21) were dose-response meta-analyses , alcohol (moderate), alcohol (heavy), total fat, saturated fat, monounsaturated fat, polyunsaturated fat, total protein, animal protein, plant protein, fruit and non-starchy vegetables, non-starchy vegetables, tomatoes, citrus fruit, alpha-carotene, beta-cryptoxanthin, lutein and zeaxanthin, dietary lycopene, total calcium (food and supplements), dietary calcium), and 67% (n=43) were high vs low meta-analyses. Of the 64 exposures, 73% (n=47) of the meta-analyses combined hazard ratios by using a random effects model, 14% (n=9) used a fixed effects model to combine hazard ratios, 11% (n=7) used a Bayesian model to combine hazard ratios, and 2% (n=1) did not provide this information. Of the 22 meta-analyses, 59% (n=13) used the Newcastle-Ottawa Scale (NOS) to assess the methodological quality of the individual studies, one of which used “the Risk Of Bias In Non-randomized Studies—of Exposures” tool in addition to the NOS to assess the methodological quality of the individual studies, and 41% (n=9) did not report a relevant methodological quality assessment instrument (n=8) or did not use a relevant tool to assess methodological quality (n=1).Except for legume fiber, the meta-analyses of each exposure included 7 (range 2–22) individual studies. Of the meta-analyses corresponding to the 64 exposures, 33% (n=10) were rated as moderate, 38% (n=11) were rated as low, and 28% (n=8) were rated as critically low. Most of the 11 low-quality meta-analyses did not meet the 3 critical items in AMSTAR 2, which were failure to state that the protocol was registered before conducting the meta-analysis, failure to conduct a comprehensive and adequate search, and failure to provide a summary of excluded publications and give reasons for exclusion. In addition to the 3 critical items mentioned above, the 8 critically low-quality meta-analyses largely failed to meet the following 2 critical items: the risk of bias of individual studies was not assessed using appropriate instruments, and the appropriate statistical methods were not used to combine hazard ratios when performing a meta-analysis [Detailed information on the methodological quality of the 29 meta-analyses assessed using ASMTAR 2 has been provided  of the associations identified in the overlapping meta-analyses were consistent in their orientation and magnitude. These associations were with the following factors: dietary fiber [n=4) of the directions and magnitudes of the associations were inconsistent, and these associations were with the following factors: red meat [Seven meta-analyses reported overlapping associations between dietary factors and the incidence of RCC –86, 88. ry fiber , 78, crury fiber , 84, vitry fiber , 85, andry fiber , 83. Howred meat –18, procred meat –18, alcored meat , 86, 88,red meat , 16. Thered meat . Detailered meat –88.P<0.05) for 18 (28%) of the 64 exposures, with 4 (67%) exposures for dietary patterns or dietary quality indices, 6 (46%) for foods with 1 (8%) that were unavailable due to missing data, 0 for beverages, 2 (29%) for alcoholic beverages, 2 (13%) for macronutrients with 4 (27%) that were unavailable due to missing data, and 4 (21%) for micronutrients. Sixty (94%) exposures in the included meta-analyses had more than 1000 cases or 20,000 participants.The SHR and level of evidence for each dietary factor are summarized in Table Table I2, small study effects, and excess significance bias could not be calculated.Table Table Table I2, small study effects, and excess significance bias could not be calculated.Table Table I2, τ2, and 95% PI are described in Additional file n=5) of I2, 8% (n=5) of τ2 and 30% (n=19) of 95% PI values could not be recalculated. In total, 67% (n=43) did not have high heterogeneity (I2<50%) for the following factors: healthy dietary pattern, unhealthy/Western pattern, drinking pattern, dietary inflammatory index, glycemic index, fish, all meat, processed meat, poultry, fruit and non-starchy vegetables, non-starchy vegetables, tomatoes, citrus fruit, coffee, beer, alcohol consumption (light), alcohol consumption (moderate), fruit fiber, cereal fiber, total fat, saturated fat, monounsaturated fat, polyunsaturated fat, total protein, animal protein, plant protein, dietary fiber, riboflavin, vitamin B6, folate, vitamin B12, methionine, choline, vitamin C, vitamin D, dietary nitrite, vitamin E, alpha-carotene, lutein and zeaxanthin, dietary lycopene, total calcium (food and supplements), dietary calcium, and calcium (supplements). For 95% PI, all exposures included null values.The n=17) included 10 or more individual studies, 31% (n=18) included between 5 and 10 individual studies, and 31% (n=18) included fewer than 3 individual studies. The contour-enhanced funnel plots corresponding to each exposure are shown  were shown in Additional file Of the 59 contour-enhanced funnel plots, 29% n=7 includep value from 7.3×10-5 to 9.18×10-10 and the evidence level for dietary fiber also increased from class IV to class III due to a decrease in p value from 7.18×10-3 to 4×10-4. Nine instances of evidence with a statistically significant difference were observed for nonsignificant results in the cohort study, but significant results in the case-control study . For processed meat, a weak level of evidence in full analysis was found, whereas the results were nonsignificant in both cohort and case-control, respectively.The results of the subgroup analyses of the associations between various dietary factors and the risk of RCC by study design are shown in Additional file The results of the subgroup analyses of the associations between various dietary factors and the risk of RCC by region are shown in Additional file p value from 6.63×10-4 to 9.71×10-3.Only vegetables maintained the same level of evidence after removing small-sized studies and low-quality studies , and nutrients, rather than a specific dietary factor, on the incidence of RCC. In our study, the evidence associating dietary factors with the incidence of RCC from available meta-analyses was summarized, and the strength and validity of this evidence were evaluated.There were 22 published meta-analyses included, with a total of 64 SHRs for various dietary factors and RCC risk. No meta-analyses were rated as high in terms of methodological quality, and approximately 1/3 were rated as moderate. Two pieces of suggestive evidence (class III) suggested that increasing vegetable and vitamin C intake might have a negative correlation with the incidence of RCC. Eight pieces of weak evidence (class IV) showed that adhering to a healthy dietary pattern was associated with a lower risk of RCC and that increased intake of cruciferous vegetables, wine, spirits, dietary fiber, vitamin E, total calcium (foods and supplements), and calcium (supplements) were negatively associated with the incidence of RCC. Seven pieces of weak evidence (class IV) showed that long-termed unhealthy pattern is positively linked with RCC incidence and that a 1-unit increase in the dietary inflammatory index, hyperglycemic index, and increased intake of all meat, processed meat, poultry, and animal protein was associated with a higher risk of RCC. The remaining dietary factors were not associated with RCC incidence.In subgroup analysis on regions, a piece of convincing evidence (class I) suggested that moderate drinking had a positive association with a low risk of developing RCC in Europe and North America. A piece of suggestive evidence (class III) suggested that increased consumption of any alcohol was positively related to a low risk of developing RCC in Europe and North America. A piece of highly suggestive evidence (class II) suggested that increased intake of cruciferous vegetables was positively linked with a low risk of RCC in North America. Two pieces of suggestive evidence (class III) suggested that increased consumption of all meat was positively related to a high risk of RCC and that increased consumption of dietary fiber was positively linked with a low risk of RCC in North America. Four pieces of suggestive evidence (class III) suggested that a healthy dietary pattern might have a negative correlation with the incidence of RCC and that an unhealthy dietary pattern and increased intake of red meat and processed meat were positively associated with a high risk of RCC in America. Two pieces of suggestive evidence (class III) suggested that increased intake of vegetables and Vitamin C is significantly associated with a lower risk of RCC in Europe.Increasing vegetable consumption was found to be inversely related to RCC risk in our study. A meta-analysis conducted by Zhang et al. found that the protective effect of vegetables was diminished after adjustment for body mass index , implyinPhytochemical extracts and compounds secreted from vegetables, such as isothiocyanates (ITC), sulforaphane (SFN), and indole-3-carbinol (I3C) , have stOur study found that increased vitamin C intake was negatively associated with the incidence of RCC. Extensive cytosine methylation abnormalities are associated with RCC tumorigenesis, and 5-hydroxymethylcytosine (5hmC) deficiency is associated with the recurrence of RCC . 5hmC dep=0.003) and asymmetric contour-enhanced funnel plots  and any drinking (class III) is more likely to be associated with a lower risk of developing RCC in Europeans and Americans compared to Asians, and a nonsignificant result that heavy drinking is not associated with RCC risk, which is consistent with what Bellocco and colleagues found . HoweverWe found a total of 18 dietary factors associated with the risk of RCC. Some results must be interpreted with caution. In case of processed meat, evidence was significant in the full analysis, whereas the results were nonsignificant in both cohort and case-control, respectively. Different study designs and the heterogeneity between studies may be one of the reasons to explain this result. To conclude, because the outcomes were nonsignificant in both study designs, the evidence of processed meat could overestimate the true effect and could thus be reconsidered. Moreover, we found that cohort studies tended to show nonsignificant results, while case-control studies tended to show significant results and higher levels of evidence, such as the 9 associations between dietary factors with RCC risk . So meta-analysis of case-control studies may have spurious associations or even seriously drive the magnitudes of the associations.p<10 −3), there was no high heterogeneity between studies, and there was no excess significance bias. Notably, both 95% PIs for exposures included null values, implying that in some cases, the effect of exposure on outcome may be nonexistent. As a result, the findings should be interpreted with caution. In addition, our study supports the statement that moderate alcohol consumption is protective, as mentioned in the European Association of Urology guidelines [There has been much interest in the topic of preventing noncommunicable diseases  by modifying dietary habits or patterns. To date, a large number of publications have reported correlations between diet and RCC incidence, including individual studies and SRoMAs, providing a large body of evidence. However, most current guidelines, including those of the European Society for Medical Oncology, the American Urological Association, and the National Comprehensive Cancer Network, do not provide recommendations on how to prevent the incidence of RCC by modifying dietary habits or patterns. Moreover, the European Association of Urology guidelines only briefly mention that specific dietary habits may influence the incidence of RCC, but the findings were inconclusive . Althougidelines , but onlIn a previous study, Papadimitriou and colleagues performed an umbrella review of evidence on the association of diet with the risk of 11 cancers (including RCC) . The preThe present study had many strengths. Our study synthesized evidence from the existing available published meta-analyses. The AMSTAR 2 instrument was employed to assess the methodological quality of the included meta-analyses. We employed credibility assessment criteria by conducting a large number of statistical tests to classify the level of evidence, as in previous umbrella reviews , 29, 30.I2, τ2, 95% PI, and excess significance bias. Therefore, the level of this part of the evidence could not be evaluated. In addition, we were unable to assess whether there was publication bias in this meta-analysis because some of the meta-analyses included too few individual studies, which prevented us from performing Egger’s test [Some limitations exist in the present study. Missing data in the included meta-analyses prevented us from calculating some metrics, such as small study effects, r’s test . During r’s test , whereasr’s test . Limitedr’s test , 180. InFundamentally, since the data in our study come from observational studies, confounding factors are inevitable . Recall Due to the rigorous items of the AMSTAR 2 instrument, the methodological quality of many of the meta-analyses was graded as low or critically low. However, inadequate reporting of meta-analyses rather than methodological flaws can also lead to a downgrading of methodological quality. Since our study included SRoMAs, the quality of the individual studies was not assessed. This should be the work of the investigators of the SRoMAs. Gray literature was not considered in this study, which may lead to some bias. Umbrella review evidence is also subject to residual confounding factors due to the type of original studies included, which were mostly observational in nature. The evidence associating legume fiber with RCC came from only one individual study , and theHeterogeneity in this study may arise from the study design, region, control selection, sex, and measurement methods. For example, the meta-analyses included in this study included case-control studies, cohort studies, and pooled studies. In addition, heterogeneity in this study may stem from the definitions of different comparator groups, as different studies set specific comparator groups, such as tertiles and quartiles of vegetable intake levels.Regarding publication bias, according to Egger’s test and contour-enhanced funnel plots, only 5 exposures in this study were potentially subject to publication bias, but 31% of the meta-analyses contained 5 to 10 individual studies, and 31% included no more than 3 individual studies , conducting a comprehensive search (including unpublished data and gray literature), providing a summary of excluded publications, giving reasons for exclusion, and selecting an appropriate quality assessment instrument to assess the risk of bias for each included individual study. Considering the potential for food interactions and the effect of confounding factors, it is recommended that studies on the effect of different dietary patterns on the incidence of RCC be conducted in the context of specific scenarios whenever possible.Overall, although a large number of SRoMAs have been published on the associations of diet with RCC incidence, none of the evidence has been classified as classes I–II in overall analysis. Increased intake of vegetables and vitamin C is inversely associated with the incidence of RCC, but the level of the evidence is class III. Moderate drinking may be beneficial for Europeans and North Americans, and cruciferous vegetables may be beneficial to North Americans. More research is needed in the future.Additional file 1: Table S1. Embase Search Strategy. Table S2. Data extraction form for included systematic reviews and individual studies. Table S3. Citation matrices for meta-analyses with overlapping associations. Table S4. List of excluded studies and reasons for their exclusion. Table S5. AMSTAR 2 quality appraisal scores. Table S6. General characteristics of meta-analyses with overlapping associations. Table S7. A. List of studies included in analysis B. List of meta-analyses with overlapping associations excluded from analysis. Table S8. Characteristics of the conducted meta-analyses and results of the recalculation, the methodological assessment (AMSTAR 2), and the level of evidence by exposure. Table S9. Summary of results of the associations on various dietary factors and risk of RCC by study design. Table S10. Summary of results of the associations on various dietary factors and risk of RCC by region. Table S11. Sensitivity analysis and the level of evidence of the association of vegetables and vitamin C with RCC.Additional file 2: Figure S1. Contour-enhanced funnel plots associating dietary patterns or diet quality indices with RCC incidence: A) healthy dietary pattern, B) unhealthy/Western pattern, C) drinking pattern, D) dietary inflammatory index, E) glycemic index, and F) glycemic load. Figure S2. Contour-enhanced funnel plots associating foods with RCC incidence: A) fish, B) vegetables, C) fruits, D) red meat, E) processed meat, and F) cruciferous vegetables. Figure S3. Contour-enhanced funnel plots associating foods with RCC incidence: A) poultry, B) all meat, C) fruit and non-starchy vegetables, D) non-starchy vegetables, E) tomatoes, and F) citrus fruit. Figure S4. Contour-enhanced funnel plots associating beverages with RCC incidence: A) sweet beverages , B) tea, C) coffee, and D) sweetened carbonated beverage. Figure S5. Contour-enhanced funnel plots associating alcoholic beverages with RCC incidence: A) beer, B) wine, C) spirits, D) alcohol (light), E) alcohol (moderate), F) alcohol (heavy), and G) alcohol (any). Figure S6. Contour-enhanced funnel plots associating macronutrients with RCC incidence: A) fruit fiber, B) vegetable fiber, C) cereal fiber, D) dietary fiber, E) Total fat, and F) saturated fat. Figure S7. Contour-enhanced funnel plots associating macronutrients with RCC incidence: A) monounsaturated fat, B) polyunsaturated fat, C) total protein, D) animal protein, and E) plant protein. Figure S8. Contour-enhanced funnel plots associating micronutrients with RCC incidence: A) riboflavin, B) vitamin B6, C) folate, D) vitamin B12, E) methionine, and F) choline. Figure S9. Contour-enhanced funnel plots associating micronutrients with RCC incidence: A) betaine, B) vitamin C, C) vitamin D, D) dietary nitrate, E) dietary nitrite, and F) vitamin E. Figure S10. Contour-enhanced funnel plots associating micronutrients with RCC incidence: A) alpha-carotene, B) beta-cryptoxanthin, C) lutein and zeaxanthin, D) dietary lycopene, E) total calcium (food and supplements), F) dietary calcium, and G) Calcium (supplements). Figure S11. After using the trim-and-fill method, contour-enhanced funnel plots for the association between A) vegetables, B) red meat, C) vegetable fiber, D) fruit and non-starchy vegetables, and E) lutein and zeaxanthin and the incidence of renal cell carcinoma were generated.