Our aim was to evaluate the accurately determine role of these two polymorphisms in CAD risk. PubMed, Embase, VIP, Wan fang and China National Knowledge Infrastructure databases were searched. The odds ratios (ORs) and 95% confidence intervals (CIs) were calculated. The trial sequential analysis (TSA) was conducted, and bioinformatics tools were employed. A total of thirty‐seven articles were obtained. For the IL‐6 rs1800795 polymorphism, 9411 CAD patients and 3161 controls were included, 4720 patients with CAD, and 5000 controls were included for the IL‐6 rs1800796 polymorphism. In the pooled analysis, significant associations were only observed for the rs1800796 polymorphism . Significant associations were detected in the Asian and Mongoloid populations and ‘more than 500’ subgroup for the rs1800795 polymorphism. TSA confirmed the true‐positive results for the rs1800796 polymorphism. The bioinformatics analysis showed that the two polymorphisms played important roles in the gene transcription. The IL‐6 rs1800796 polymorphism is associated with an increased susceptibility to CAD and is a risk factor for CAD. The IL‐6 rs1800795 polymorphism is associated with an increased risk of CAD in Asians, particularly in Chinese, and a decreased risk of CAD in an African population is remarkably observed.Studies examining the associations between the  Several previous studies have been conducted in an attempt to draw significant conclusions, but the limitations in sample size and potential false‐positive results caused by systematic errors may bias the results. We therefore performed a study to more accurately determine associations between IL‐6 polymorphisms and CAD risk; in addition, the bioinformatics analysis was conducted to explore the potential molecular mechanism.Inflammation plays a key role in the pathophysiology of CAD by promoting the development of atherosclerosis.22.1interleukin‐6 (IL‐6) gene and coronary artery disease. The terms ‘coronary artery disease’, ‘coronary heart disease’, ‘myocardial infarction’, ‘CAD’, ‘heart disease’, ‘interleukin‐6’, ‘IL‐6’, ‘polymorphism’, ‘variant’ and ‘polymorphisms’ were used. The citations of review articles and all eligible studies were also browsed for additional potentially relevant study data. In addition, the language of the published studies was restricted to English.A comprehensive document retrieval procedure was conducted to identify for all relevant studies published prior to October 2019. PubMed, Embase, VIP, Wan fang and China National Knowledge Infrastructure databases were thoroughly searched by the first three investigators to identify potential studies examining the associations between polymorphisms in the 2.2IL‐6 polymorphisms and coronary artery disease; (b) coronary artery disease was defined as 50% stenosis in the left main coronary artery, or multiple significant (≥70% stenosis) in more than one coronary arteryFor inclusion in our analysis, studies must have met the following inclusion criteria: (a) evaluation of the relationship between the 2.3IL6 gene in patients and controls. Disagreement was settled by rechecking the data or discussion with a third author.For all included studies, the first two authors independently extracted the following data using a standardized form: first author's last name, year of publication, study country, study region, age and body mass index (BMI), source of the control population, genotyping method, sample size and genotype frequency of polymorphisms in the 2.4The quality of the included studies was independently assessed by all the authors according to a set of criteria that were modified based on the Newcastle‐Ottawa quality assessment scale Table .2.5P < .05 was considered a significant departure from HWE. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to evaluate the strength of the associations between IL‐6 gene polymorphisms and coronary artery disease risk. Pooled ORs were calculated for the allelic model (IL‐6 rs1800795: C versus G and IL‐6 rs1800796: G versus C), recessive model (IL‐6 rs1800795: CC versus CG + GG and IL‐6 rs1800796: GG versus GC+CC), dominant model (IL‐6 rs1800795: CC+CG versus GG and IL‐6 rs1800796: GG+GC versus CC), heterozygote model (IL‐6 rs1800795: CG versus GG and IL‐6 rs1800796: GC versus CC) and homozygote model (IL‐6 rs1800795: CC versus GG and IL‐6 rs1800796: GG versus CC), respectively. Heterogeneity was evaluated using the Q statistic (significance level of P < .1) and I2 statistic (greater than 50% as evidence of a significant inconsistency). Heterogeneity between studies was evaluated with the I2 test, and a higher I2 values indicated higher levels of heterogeneity . In the heterogeneity evaluation, the fixed‐effects model was used when I2 < 50%, a random‐effects model was used if I2 = 50% to 90%, and the studies were not pooled if I2 > 90%. A sensitivity analysis was performed to detect heterogeneity by sequentially omitting each study. Additionally, analyses were performed in subgroups stratified by accordance with HWE, region, ethnicity, source of controls and sample size. The potential for publication bias was assessed with Begg's funnel plot and Egger's test. We applied the Bonferroni method,Hardy‐Weinberg equilibrium (HWE) was tested in control groups from each study using the chi‐squared test, and 2.6Systematic bias and random errors are inevitable when conducting a meta‐analysis because of the sparse data and repeated significance testing; moreover, trials with low methodological quality, publication bias and a small sample size may generate a false‐positive result. Trial sequential analysis (TSA) is an approach that provides the required amount of information (number of samples) and further reveals potentially false‐positive results in a meta‐analysis.2.7http://asia.ensembl.org/index.html). SNPinfo is an important bioinformatics analysis tool that predicts SNP function. The SNPinfo database can help researches specify genes or linkage regions and select SNPs based on GWAS results, calculate linkage disequilibrium (LD) and predict functional characteristics of both coding and non‐coding SNPs (https://snpinfo.niehs.nih.gov/).Ensembl is a genome browser for vertebrate genomes that supports research in comparative genomics, evolution, sequence variation and transcriptional regulation, and this database provides the genomic context, genes and regulatory elements, flanking sequence, population genetics, phenotype data, sample genotypes, linkage disequilibrium and phylogenetic context of a single nucleotide polymorphism , dominant genetic model , heterogeneity genetic model , recessive genetic model .The main results of our analysis and the heterogeneity test of the associations between 3.3IL‐6 polymorphisms and CAD risk. Table Subgroup analyses were introduced to identify the source of heterogeneity and further reveal additional information about the associations between IL‐6 rs1800796 polymorphism, and all five genetic models indicated strong associations with an increased OR compared with the pooled OR: allelic , dominant , recessive , heterozygote  and homozygote  genetic models.For the subgroup in accordance with HWE, significant associations were only detected for the P = 2*10−4), recessive   and homozygote  genetic models. In addition, decreased risks for the African population were remarkably identified in the allelic  and recessive  genetic models. Regarding the rs1800796 polymorphism, all five genetic models suggested a strong relationship with CAD risk in Asian volunteers . Along with region as a geographic factor, ethnicity is also an important factor. In the Caucasian population, no association was observed for the rs1800795 polymorphism, but significant associations were widely observed with the rs1800796 polymorphism in all five genetic models . Regarding the Mongoloid population, significant associations were observed in the dominant , recessive , heterozygote  and homozygote  genetic models that implied a strong association between CAD risk and the IL6 rs1800795 polymorphism.In the analysis of the rs1800795 polymorphism stratified by region, significant associations with reduced heterogeneity in the Asian population were observed in the dominant  Figure , heteroz1 Figure ; heterozP = .009) and the rs1800796 polymorphism .In the subgroup analysis stratified by source of controls, significant associations were observed between CAD risk in the hospital‐based population and the rs1800795 polymorphism ; for the rs1800796 polymorphism, a significant association was only observed in the allelic genetic model . We discovered that a larger sample produced more significant associations with the two polymorphisms.We stratified studies into three subgroups by sample size based on the modified quality scale score  to evaluate the effect of sample size on the associations between the two polymorphisms and CAD risk. In the greater than 500 subgroup, significant associations were observed between the rs1800795 polymorphism and CAD risk , but also cross the O’ Brien Fleming boundaries, indicating that the significance level of our study was a true‐positive result and the previously reported negative association was due to a lower number of volunteers.A previous meta‐analysis of the associations between 3.7Based on the genomic context obtained from the Ensembl database, we constructed the summary genetic diagram for the rs1800795 and rs1800796 polymorphisms Figure . The two4IL‐6 gene were analysed for associations with CAD risk. The two common polymorphisms have been extensively studied in depth over the past few decades, providing sufficient enough data for a subgroup analysis designed to discover potential intriguing associations. Moreover, the two polymorphisms are located in the promoter region of the IL‐6 gene, and may influence the expression of the IL‐6 gene, and result in susceptibility to CAD. Several meta‐analyses have been conducted to explore the associations between the two polymorphisms and CAD risk, but the results were inconsistent. Significant associations between the IL‐6 rs1800795 polymorphism and CAD risk were reported in some meta‐analyses,IL‐6 rs1800796 polymorphism. The most recent studies by Song et alIL6 rs1800795 and rs1800796 polymorphisms and CAD risk. Therefore, we chose these two common polymorphisms in the IL6 gene to analyse the potential CAD risk.In our study, two polymorphisms (rs1800795 and rs1800796) in the IL‐6 rs1800795 polymorphism and CAD risk with high heterogeneity was observed in the pooled results. Hence, we employed a detailed subgroup analysis to determine the potential sources of heterogeneity and associations. For the subgroup analysis stratified by region, significant associations with reduced heterogeneity were observed in the Asian population, which indicated an increased CAD risk for the rs1800795 polymorphism. Interesting results emerged in the analysis of the African subgroup and the decreased CAD risk for the mutant C allele and CC genotype were observed compared with the wild G allele and GC+CC genotypes, respectively. However, only three studies were included in the analysis of the African subgroup, and thus, we are sceptical about the conclusions and further studies are required. Similar findings were obtained for the Asian population. When stratified by ethnicity, the increased CAD risks for the rs1800795 polymorphism were observed in the Mongoloid population. The studies in the Mongoloid subgroup were performed in China, indicating that Chinese patients carrying the rs1800795 polymorphism would exhibit an increased risk of CAD. Moreover, high heterogeneity was significantly reduced when volunteers were stratified by region and ethnicity, indicating that these two factors are potential sources of high heterogeneity for the rs1800795 polymorphism. Sample size plays an important role in interpreting the conclusions of a case‐control study; thus, we conducted an analysis of subgroups stratified by sample size and discovered the increased CAD risk for the rs1800795 polymorphism in the larger sample size group. Therefore, if additional well‐designed studies are conducted, a lager sample size is required. The IL‐6 rs1800796 polymorphism is associated with an increased risk of CAD. Extensive associations with an increased risk of CAD were observed in the pooled analysis and subgroup analyses. In addition, the trial sequential analysis confirmed the true‐positive results for the rs1800796 polymorphism, indicating that carriers of the rs1800796 mutant G allele are likely predisposed to CAD.No association between the IL‐6 gene may also be an answer to this question. We conducted a bioinformatics analysis to predict the potential molecular mechanism. The two polymorphisms are located in the promoter region of the IL‐6 gene, implying that the underlying mechanism occurs at the transcriptional level. The analysis of the SNPinfo database revealed potential transcription factor binding sites in the two polymorphisms, indicating the ability of these two polymorphisms to alter the expression of the IL‐6 gene. In addition, an analysis of the sequence and secondary structure was performed using the RNAfold web server. The minimum free energy (MFE) and the free energy of the thermodynamic ensemble of the mutant alleles of the rs1800795 and rs1800796 polymorphisms were reduced compared with the wild alleles. The principle of minimum energy states that for a closed system, with constant external parameters and entropy, the internal energy will decrease and approach a minimum value at equilibrium.IL‐6 gene. However, a bioinformatics prediction is not sufficient, and further fundamental research on the effect of the two polymorphisms on the transcription of the IL6 gene is needed.Coronary atherosclerosis is the main pathophysiological process in coronary artery disease, and inflammation plays a predominant role in atherosclerosis. As an important pro‐inflammatory cytokine, IL‐6 has been proven to be an independent risk factor for coronary artery disease and is expressed at relatively high levels in human atherosclerotic plaques.IL‐6 rs1800795 polymorphism, and thus, the results should be interpreted with caution. Third, only two common SNPs were evaluated in our study and other relevant SNPs in the IL‐6 gene that are unknown or understudied may also have potential associations with CAD risk. Forth, the distance between the two common polymorphisms is relatively close, and the potential interactions between the two polymorphisms or other unknown polymorphisms need to be studied. In addition, the potential influence of environmental factors on genotype‐CAD associations is worth considering.Several limitations existed in our study. First, only English and Chinese articles were included as a language restriction, which may bias the results. Second, the number of included studies was relatively small in some subgroups, such as the African population in the subgroup analysis of the IL‐6 rs1800796 polymorphism is associated with an increased susceptibility to CAD and is a risk factor for CAD. In addition, the IL‐6 rs1800795 polymorphism is associated with an increased risk of CAD in Asian, particularly in Chinese volunteers. Remarkably, a decreased risk of CAD was observed in the African population.In conclusion, the The authors confirm that there are no conflicts of interest.Shuai Lu, Ya Wang and Zhaohui Wang wrote the main manuscript text. Yijun Wang and Shuai Lu prepared Figures 1 and 2. Wenhan Ma and Jing Hu prepared Figures 3 and 4. Shuangye Liu, Di Wu and Xiaohui Zeng draw the Figures 5 and 6. Yijun Wang made the Table 1. Shuai Lu, Ya Wang and Yijun Wang forged the Table 2. Guo Yu performed the statistical analysis of evaluating the strengths of associations between these two polymorphisms and CAD risk. Zhaohui Wang hosted the meeting addressing the disagreements. All authors reviewed and revised the manuscript.Table S1Click here for additional data file.Table S2Click here for additional data file.


Although neutrophils have been linked to the progression of cancer, uncertainty exists around their association with cancer outcomes, depending on the site, outcome and treatments considered. We aimed to evaluate the strength and validity of evidence on the association between either the neutrophil to lymphocyte ratio (NLR) or tumour-associated neutrophils (TAN) and cancer prognosis.We searched MEDLINE, Embase and Cochrane Database of Systematic Reviews from inception to 29 May 2020 for systematic reviews and meta-analyses of observational studies on neutrophil counts (here NLR or TAN) and specific cancer outcomes related to disease progression or survival. The available evidence was graded as strong, highly suggestive, suggestive, weak or uncertain through the application of pre-set GRADE criteria.N = 9). When assessed for magnitude of effect, significance and bias related to heterogeneity and small study effects, 18 (9%) associations between NLR and outcomes in composite cancer endpoints , cancers treated with immunotherapy and some site specific cancers  were supported by strong evidence.A total of 204 meta-analyses from 86 studies investigating the association between either NLR or TAN and cancer outcomes met the criteria for inclusion. All but one meta-analyses found a hazard ratio (HR) which increased risk (HR > 1). We did not find sufficient meta-analyses to evaluate TAN and cancer outcomes  meta-analyses presented strong or highly suggestive evidence. Although the NLR and TAN hold clinical promise in their association with poor cancer prognosis, further research is required to provide robust evidence, assess causality and test clinical utility.CRD42017069131.PROSPERO  Cancer is the second leading cause of mortality worldwide , contribThe NLR has emerged as a potential biomarker of cancer prognosis and is of particular clinical interest due to its accessibility and the ease of calculating the ratio from patients’ routine blood cell counts . The NLRIt is currently unclear how the association between NLR and poor prognosis varies depending on the site of cancer or the treatment considered. Umbrella reviews allow for the analysis of such broad subject areas to examine the strength and credibility of associations using the results of published systematic reviews and meta-analyses , 15. UmbSearches were conducted in MEDLINE, Embase and the Cochrane Database . However, meta-analyses which assessed multiple cancers in a single analysis based on a shared feature were included and classified as “composite cancer endpoints”. Meta-analyses were also excluded if they did not provide sufficient detail for replication, such as the hazard ratio (HR), 95% confidence interval and total sample size of each individual study included in a meta-analysis. If a single systematic review included multiple meta-analyses, all meta-analyses were individually assessed for eligibility.When more than one meta-analysis was identified for a single association at a specific site, they were assessed for concordance in the direction, magnitude and significance of their effect estimates. If the duplicate meta-analyses identified agreed in significance, magnitude and direction of effect, the meta-analysis with the greatest number of component studies was included. Where disagreement was found, both duplicate meta-analyses were excluded unless the disagreement arose from an underpowered meta-analysis (less than five component studies), in which case the meta-analysis with the greatest number of component studies was used.Data extraction forms were generated to record information from each meta-analysis and the included individual studies. First author, year of publication, outcome measure, biomarker and cancer diagnosis were extracted from each meta-analysis. For each included individual study within a meta-analysis, the first author, year of publication, total population, epidemiological design, HR and 95% confidence interval were extracted, along with analysis method and NLR cut-off where available.The weighted inverse variance method was used with restricted maximum likelihood (REML) estimations to reproduce all included meta-analyses in R with the “meta” package  and “metEach included meta-analysis was reproduced to yield both fixed and random effects estimates. Reproduced random or fixed effect estimates which did not match the results of the original review results were assessed for absolute and percent difference. Meta-analyses with a difference in HR of 0.01 were attributed to rounding errors. Studies with larger discrepancies were investigated to determine the source of disagreement. Where there were issues with reproducibility, the calculated values of the random effects model were used to assess the evidence for the association.Tau) [I2 statistic [P < 0.10 [I2 statistic was also used to quantify the percentage of variation which can be attributed to heterogeneity [I2 value were included to evaluate the uncertainty around estimates of heterogeneity [We calculated 95% prediction intervals (PI) in order to assess the impact of uncertainty around the summary effect size estimate and between-study variance (Tau) . PredictTau) . In studtatistic . We consP < 0.10 . Due to ogeneity . The 95%ogeneity . Howeverogeneity , were prP < 0.05).To further explore heterogeneity and determine the impact of adjustment with additional prognostic factors, sub-group analyses of the statistical models used in each study were conducted. Meta-analyses which reported the analysis method (univariate or multivariate) of their component studies were included. The Q test for subgroup differences was used to test for significant differences in the random effects model effect size between subgroups in each meta-analysis  was used to indicate presence of small study effects which could reflect publication and other selective reporting biases. Further assessment was carried out to determine if the summary effect size estimate of each meta-analysis was greater than the point estimate of the largest included study, indicating potential small study effects [Small study effects and funnel plot asymmetry were quantified through the arcsine-Thompson test described by Rücker et al.  and the  effects . A meta-P < 0.10) [The test for excess significance (TES) was used to determine if the number of observed significant results differed significantly from the expected number, indicating reporting bias . TES res < 0.10) . The exp < 0.10) , 30. The < 0.10) , using tc that the true effect size is in the opposite direction of effect of the observed point estimate [Credibility ceilings were utilised for sensitivity analysis and to test the robustness of meta-analyses, considering studies of biomarkers often underestimate biases due to large sample sizes and observational study design , 33. Creestimate , 32. The. In order for an association to be considered strong, the meta-analysis must yield a P value of less than 10− 6 in the random effects model [P < 0.05 in the largest included study, find no heterogeneity (P > 0.10) through the Q test, detect less than 50% variance due to heterogeneity through the I2 statistic, yield a prediction interval excluding the null value (HR = 1), display no evidence of small study effects or excess significance, and the association must maintain significance at P < 0.05 with the application of a credibility ceiling of 10%. The number of studies in each meta-analysis was also included as eligibility criterion for strong evidence since a sample size greater than three is required for reliable assessment of heterogeneity and small study effects [I2 and Cochran’s Q test criteria were removed for associations classified as strong. Additionally, we applied a simplified method to rank associations based only on their effect size estimate (HR) and precision  to aid interpretation.Associations between neutrophil counts (here NLR or TAN) and cancer prognosis were categorised into strong, highly suggestive, suggestive, weak or uncertain through assessment of the strength and validity of the evidence for each meta-analysis, according to pre-defined Grading of Recommendations, Assessment, Development and Evaluations (GRADE) criteria outlined in Additional File ts model , include effects , 36, 37. effects . To asseStudies with meta-analyses categorised as providing either highly suggestive or strong evidence underwent quality assessment through AMSTAR 2, a tool for assessing the methodological quality of systematic reviews for both health interventions and observational research , 40. FurStatistical analyses were carried out in R , includiFunders had no role in data collection, analysis, interpretation or writing of the report. All authors had access to all the data in the study.No patients were involved in development of our umbrella review design nor were they asked to advise on interpretation. No ethical approval was required for this review since it relied entirely on anonymised, published data.N = 195) or TAN (N = 9) categorically as either a high or low value. Included meta-analyses summarised effect size estimates from 1978 individual studies, with OS as the most frequently assessed outcome (N = 90). In 135 meta-analyses (66%), total sample size exceeded 1000 individuals and each meta-analysis had a median of seven studies. However, 134 meta-analyses (66%) included less than ten studies and 25 (12%) included only two studies. The characteristics of included meta-analyses are summarised in Additional File The 140 published articles meeting the criteria for inclusion contained 517 meta-analyses  were significant with random effects. At a more stringent threshold of P < 10− 6, the number of statistically significant meta-analyses for random effects dropped to 93 (46%)  assessed NLR as a biomarker of poor prognosis in melanoma, neurologic, gynaecologic, pancreatic, gastrointestinal and oesophageal, colorectal, hepatocellular and biliary, respiratory and oral, urinary, head and neck, soft tissue sarcoma (STS), treatment with immunotherapy, and composite cancer endpoints. TAN  were assessed as a biomarker of poor prognosis in three of the 93 meta-analyses (3%), including urinary and composite cancer endpoints.Using a threshold of P < 0.05. However, 42 (88%) of these meta-analyses still had a statistically significant summary random effects estimate. In three meta-analyses [In 48 meta-analyses 24%), the largest component study was not statistically significant at analyses , 70, 94,4%, the lIn order to determine the impact of study size on the magnitude of the summary effect size, random effects estimates were plotted against the inverse variance of the pooled effect size from each meta-analysis. When compared to meta-analyses with large variances, those with smaller variances produced more conservative estimates, displaying a smaller range of HR estimates and a slight tendency toward a null value (HR = 1). Meta-analyses with large variance displayed a wide range of random effects HR and included an increased number of HR estimates greater than two Fig. b.In 87 of the 204 included meta-analyses (43%), the pooled effect size was reproduced with an absolute difference between the calculated and reported HR outside of the range which can be attributed to rounding errors > 0.01). Twenty-eight of these 87 meta-analyses were within 2% of the reported HR, 35 were between 2% and 5% of the reported HR, 14 were between 5% and 10% and ten meta-analyses reported an HR with over a 10% difference from the calculated HR  meta-analyses which had included only two individual studies . In general, adjustment of HRs resulted in minor modification in the strength of association with no consistent pattern. In most cases, summary estimates obtained with the random effects model showed slightly stronger or weaker associations with no change in the direction of effect and no major differences in the width of the 95% CI. However, we note that the majority of studies had 10 or fewer component studies. In studies where both univariate and multivariate meta-analyses included at least five component studies, 95% PIs tended to be wider for multivariate studies, with lower bounds close to the null.In the subgroup analyses of the 48 meta-analyses 24%) which reported on the analysis methods utilised by component studies, a significant difference was found between the multivariate and univariate groups in only six 13%) meta-analyses , ethnicity and current medications were not typically reported and could modify the estimates.2 in many cases, likely due to small sample sizes [2 value of 0%. In 15 (20%) of the remaining 74 meta-analyses, the addition of NLR cut-off values accounted for 100% of the heterogeneity, with a mean and standard deviation of 50.15 ± 39.94%.156 meta-analyses (76%) reported sufficient information on NLR cut-offs to undergo meta-regression analysis  included the recommended threshold of ten or more component studies to be considered reliable . Of thesP-values (P < 0.10), indicating potential small study effects. The presence of small study effects was also assessed through comparison of the random effects model effect estimate and the effect estimate of the largest component study. In 68 meta-analyses (33%), the summary effects estimate from the largest component study was larger than the point estimate of the model and were considered to have evidence of small study effects. Taking both criteria into account, 136 (67%) of the 204 included meta-analyses were judged to have evidence of small study effects  included two or more studies and were eligible for further assessment through the arcsine-Thompson test for publication bias . Eighty-Forty-two meta-analyses 21%) showed evidence of excess significance bias according to the TES when the effect size of the largest component study was utilised as an estimate of true effect size   and 139 (68%), 84 (41%) and 51 (25%) maintained significance at ceilings of 10%, 15% and 20%, respectively. All of the meta-analyses remained consistent in direction of effect (HR > 1) up to a ceiling of 15% and three (1.5%) yielded an effect estimate in the opposite direction (HR < 1) with a ceiling of 20%. The P < 0.05. The remaining 188 meta-analyses (92%) provided at least weak evidence of an association (P < 0.05) , OS and PFS in nasopharyngeal cancer , OS in castration-resistant prostate cancer , RFS in bladder cancer with urothelial transurethral resection (UTR) , OS in endometrial cancer  and DFS in soft tissue sarcoma (STS)  and the other 11 were upgraded from highly suggestive , including DFS in composite cancer endpoints , OS in advanced cancer with anti-vascular endothelial growth factor receptors (VEGFR) , OS in cancer with immune checkpoint inhibitors (ICI) , OS in gastric cancer with surgical resection (SR) , OS in colorectal liver metastasis (CLM) , OS in CLM with SR , OS in breast cancer , OS in renal cancer with tyrosine kinase inhibitors (TKIs) , OS in melanoma with ICI , OS in non-small cell lung cancer (NSCLC) with PD-1 inhibitors  and OS in breast cancer with SR .Strong evidence was presented in 18 meta-analyses (9%) for associations between NLR and poor cancer prognosis. Seven of these associations met the grade criteria for strong evidence, including PFS in non-muscle-invasive bladder cancer (N = 26), followed by PFS (N = 7), DFS (N = 5) and RFS (N = 4).Forty-two meta-analyses (21%) presented associations supported by highly suggestive evidence, including associations between increased NLR and poor prognosis in composite cancer endpoints, cancers treated with immunotherapy, gastric, colorectal, CLM, pancreatic, gynaecologic, breast, hepatocellular, biliary, NSCLC, lung, head and neck, oral, renal, advanced renal cancer, upper urinary and bladder, STS and bladder. The most commonly assessed outcome for highly suggestive associations was OS (N = 50) or TAN (N = 1) and poor cancer prognosis, and 77 meta-analyses (38%) provided weak evidence for an association between high NLR or TAN and poor cancer prognosis. The association between intratumoural neutrophils and overall survival in composite cancer endpoints was classified as suggestive, but there was weak evidence supporting associations with peritumoural neutrophils or stromal neutrophils. Details of the grading and upgrading for each meta-analysis are included in Additional File Fifty-one meta-analyses (25%) provided suggestive evidence for an association between high NLR . Although GRADE criteria are limited by rigid thresholds, these analyses suggest they are relatively robust and considerably improve the classification of evidence for quality and strength of recommendations.The sensitivity analysis of GRADE criteria resulted in only one association between NLR and OS in gastric cancer being reclassified from highly suggestive to strong when N = 26) or strong evidence (N = 12) arose from 30 individually published studies. Of the 26 highly suggestive meta-analyses, four were from studies ranked as critically low quality (15%), five as low quality (19%), seventeen as moderate quality (65%) and one as high quality (4%). The twelve meta-analyses categorised as providing strong evidence were ranked as moderate quality , low quality  and critically low quality   to be at low risk of bias . However, a cautious interpretation is required due to the presence of heterogeneity and small study effect biases. Associations supported by strong evidence included elevated NLR in urinary (prostate and non-muscle-invasive bladder) and nasopharyngeal cancers, amongst others. Future research should assess the association between the NLR and prognosis in oral and respiratory cancers by environmental exposures, such as smoking status, to ensure these do not confound associations.Previous studies have documented the prognostic role of neutrophils, particularly the NLR, and their link with poor outcomes for many cancer sites . We founCLM represents a unique case where metastasis has already occurred and may present a link between elevated NLR and poor prognosis in metastasised cancers. It is interesting to note that the associations between NLR and OS and PFS in colorectal cancer also included patients with different stages of metastasis according to the tumour node metastasis (TNM) system. These associations were supported by highly suggestive and suggestive evidence, respectively, and emerged from a study of moderate quality. To determine the impact of advanced cancer stages, metastasis and subtypes of cancer on the association between the NLR and cancer outcomes, future studies should consider these additional factors carefully when assessing the prognostic potential of the NLR.Although previous studies suggest a link between TAN and the progression of cancer in the tumour microenvironment , 13, 130A key strength of this study comes from the use of umbrella review methodology which only includes meta-analyses as evidence for quantitative data analyses . The useAlthough there are considerable differences between the included meta-analyses in terms of cancer site, stage and treatment, all but one of the HR estimates reported in these meta-analyses were in the same direction of effect. This finding suggests consistency in the relationship between neutrophil biomarkers and poor outcomes across a wide spectrum of cancer diagnoses.Only 39% of the identified meta-analyses were eligible for inclusion and 16% of the identified meta-analyses were excluded because they did not include sufficient data to be reproduced. Furthermore, we were unable to reproduce 43% of the included studies within 0.01 of the reported HR, highlighting issues with transparency and reproducibility of findings in epidemiologic research . UmbrellUmbrella reviews are reliant on the quality of the included meta-analyses. This methodological limitation is of concern since 42% of the studies which yielded meta-analyses with highly suggestive or strong evidence were ranked as low or critically low quality by AMSTAR 2 . Appendix J: Effect size upgrading of GRADE. Appendix K. Sensitivity analysis of GRADE. Appendix L. Ranking based on precision versus effect size. Appendix M. AMSTAR 2 Quality Rating of Meta-analyses with Highly Suggestive or Strong Evidence. Appendix N. Assessment of quality of component studies. Appendix O: Sensitivity analysis of overall survival in rectal cancer.Additional file 2. Supplementary Figs. 1–205.Additional file 3. Supplementary Table 1.


Background/Aim: Several observational studies evaluated the links between serum monounsaturated fatty acids (MUFAs) and cardiovascular events with controversial results. In the present study, Mendelian randomization (MR) analysis was applied to obtain unconfounded estimates of the causal associations of genetically determined serum MUFAs with coronary heart disease (CHD), myocardial infarction (MI), cardioembolic stroke (CS), and ischemic stroke (IS).Methods: Four MUFAs were studied . Data from the largest genome-wide association studies on MUFAs, CHD, MI, and stroke were analyzed. Inverse variance weighted method (IVW), weighted median (WM)-based method, MR-Egger, as well as MR-pleiotropy residual sum and outlier were applied. To rule out the impact of single-nucleotide polymorphism (SNP), the leave-one-out method was also performed.Results: Genetically higher-serum 10-heptadecenoate levels did not affect the risk of CHD , MI , CS , and IS . Similarly, no significant associations were observed for myristoleic acid , oleic acid , and palmitoleic acid . A low likelihood of heterogeneity and pleiotropy was reported, and the observed associations were not driven by single SNPs.Conclusions: In the present MR analysis, serum MUFA levels were not associated with the risk of CHD, MI, CS, and IS. Further research, evaluating more MUFAs, is required to elucidate the links between MUFAs and CVD to contribute to health policy decisions in reducing CVD risk. Major snd trans . MUFAs and trans .It is widely known that fatty acid metabolism and composition can be altered during diseases, resulting in beneficial , 5 or advia promoting the inflammation process  synthesis , a CVD r process .The activity of SCD1 is enhanced in various cardiometabolic diseases, including diabetes, obesity, and atherosclerosis . The incOverall, there is conflicting and limited evidence in relation to the links between serum MUFAs and CVD. The aim of the present study was to evaluate the causal associations of genetically determined serum MUFAs with coronary heart disease (CHD), myocardial infarction (MI), cardioembolic stroke (CS), and ischemic stroke (IS) by applying Mendelian randomization (MR) analysis on data from the largest genome-wide association studies (GWAS) on MUFAs.A two-sample MR study design was used, in which summary statistics from different GWAS were analyzed for the exposures (serum MUFAs) and outcomes , to estimate the effects of exposure on outcome . Essentir2 = 0.8. To minimize bias in effect estimates induced by the correlation between SNPs, we restricted our genetic instrument to independent SNPs not in LD (p = 0.0001). We refer to a set of SNPs that proxy serum MUFAs as “genetic instruments.”We retrieved summary data for the associations between single-nucleotide polymorphisms (SNPs) and four circulating MUFAs  from the GWAS  . Genotypn ≤ 76,014)–control  study based on a meta-analysis, with the use of double genomic control correction, of the CARDIoGRAMplusC4D 1000 Genomes case–control  study, the UK Biobank SOFT CAD study , and two small case –control  studies from Germany and Greece  method. To address the potential effect of pleiotropic variants on the final effect estimate, we performed sensitivity analysis, including weighted median (WM) and MR-Egger. The sensitivity analysis was conducted using the leave-one-out method to identify instruments that might drive the MR results. The WM estimate provides correct estimates as long as SNPs accounting for ≥50% of the weight are valid instruments. Inverse variance is used to weight the variants and bootstrapping is applied to estimate the confidence intervals (CIs) . MR-EggeMR-Egger . FurtherMR-Egger . The anaMR-Egger .Q′ heterogeneity statistic  tests were used for sensitivity analysis . MR-EggeMR-robust adjusted profile score (RAPS) was also applied. This method can correct for pleiotropy using robust adjusted profile scores. We consider, as results, causal estimates that agreed in direction and magnitude across MR methods, passed nominal significance in IVW MR, and did not show evidence of bias from horizontal pleiotropy using heterogeneity tests. All analyses were performed using the R software .The present analysis uses published or publicly available summary data. No original data were collected for this study. Ethical approval for each of the studies included in the present analysis can be found in the original publications (including informed consent from each participant). The study conforms to the ethical guidelines of the 1975 Declaration of Helsinki.The instrument associations for serum MUFA levels are shown in the p = 0.185, p = 0.066, p = 0.945, p = 0.767, p = 0.972; for MI: IVW = Beta: 0.041, p = 0.887; for CS: IVW = Beta: 0.881, p = 0.349; for IS: IVW = Beta: 0.162, p = 0.727, p = 0.6334; for MI: IVW = Beta: −0.119, p = 0.8125; for CS: IVW = Beta: 1.059, p = 0.4419; for IS: IVW = Beta: 0.008491, p = 0.9901, p = 0.782; for MI: IVW = Beta: −0.01255, p = 0.9648; for CS: IVW = Beta: 1.042, p = 0.3687; for IS: IVW = Beta: −0.1862, p = 0.7137, Genetically higher-serum 10-heptadecenoate levels had no significant effect on the risk of CHD . Furthermore, MR-PRESSO analysis did not indicate any outliers for all estimates. The horizontal pleiotropy test, with very negligible Egger regression intercept, also showed a low likelihood of pleiotropy for all of our estimations . The results of the MR-RAPS were identical with the IVW estimates, highlighting again a low chance of pleiotropy. The leave-one-out method demonstrated that the associations were not driven by single SNPs.Heterogeneity results and pleiotropy bias are also shown in The present MR analysis showed that genetically higher-serum MUFA  levels did not affect the risk of CHD, MI, CS, and IS. These associations were not driven by single SNPs, and there was a low chance of heterogeneity and pleiotropy.Heptadecanoic acid was inversely associated with the risk for MI in 385 women in a population-based cohort ; in a multivariate analysis, however, this link became non-significant . Furtherp = 0.008) among 6,568 participants of the Multi-Ethnic Study of Atherosclerosis (p = 0.03) in a case–control study (n = 297) (p = 0.03) among 1,245 participants from the Three-City Study followed up for a median of 5.25 years  . Furthern = 297) . In this25 years .p = 0.03) in a case–control study (n = 297) ; p = 0.005] in 3,870 participants from the Minneapolis field center of the Atherosclerosis Risk in Communities study  found no association between trans-palmitoleic acid, CHD, and stroke  . In contn = 297) . In thisn = 297) . Furtheres study . Similard stroke . Trans-pd stroke . Finallyd stroke .In conclusion, the present MR analysis showed that serum MUFA levels did not correlate with the risk of CHD, MI, CS, and IS. Further studies, investigating also other MUFAs, are needed to clarify the links between MUFAs, CHD, MI, CS, and IS. This research is clinically important for health policy decisions in reducing CVD risk.All datasets presented in this study are included in the article/All authors contributed to the design, interpretation, and writing of the paper.NK has given talks, attended conferences, and participated in trials sponsored by Astra Zeneca, Bausch Health, Boehringer Ingelheim, Elpen, Menarini, Mylan, NovoNordisk, Sanofi, and Servier. The remaining 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.


The incidence of osteoporotic fractures has increased rapidly, and because of the poor prognosis and high mortality associated with osteoporotic fractures, they remain a prospective research area globally. One way to reduce their incidence is to investigate their intervention risk factors in the elderly. Hence, this study explores the correlation between serum 25-hydroxyvitamin D [25(OH)D] levels and osteoporotic fractures in elderly patients through a meta-analysis.We conducted our literature search mainly in PubMed and Embase for identifying studies that investigated the relationship between serum 25(OH)D levels and the risk for osteoporotic fractures. We performed categorical analysis, heterogeneity checks, publication bias analysis, and subgroup analyses.In total, 20 studies were included, of which 4 were case-cohort studies and 16 were cohort studies. A total of 41,738 patients from 20 studies were included in the meta-analysis, of which 5916 had fractures, including 3237 hip fractures. By combining the lowest and highest categories of relative risks (RRs) and 95% confidence intervals (CIs), it was suggested that lower serum 25-hydroxyvitamin D levels may be a risk factor for fractures. RR (95% CI) for total and hip fractures were 1.11  and 0.89  after adjustments.Our study showed that compared to low serum 25(OH)D levels, high serum 25(OH)D levels reduce the risk of hip fractures in the patients aged 60 years or older. In contrast, serum 25(OH)D has no significant relationship with total fracture risk. As the growth of population worldwide, the incidence of osteoporotic fractures is rising rapidly as well . The lifSerum 25-hydroxyvitamin D [25(OH)D] is recognized as the main circulating form of vitamin D, which accurately indicates the vitamin D concentration in the body. It reflects the nutritional status of vitamin D. Bone mineral density (BMD), bone size (relative to body size), and the bone strength impacted by the level of serum 25(OH)D , 4. PrevHowever, it is worth noting that due to the different bone turnover rates, osteoporotic fractures in the elderly are of two types: perimenopausal and senile osteoporotic fractures . When thThere is no meta-analysis that presents the association between serum 25(OH)D and the risk for osteoporotic fractures in the elderly older than 60 years. Our study answers the following question for serum 25(OH)D and osteoporotic fractures: What is the relationship between serum 25(OH)D and hip and total fractures in elderly older than 60 years?We performed a meta-analysis of the available literature according to the PRISMA statement guidelines . EthicalWe searched PubMed and EMBASE systematically using the following keywords: (1) “serum 25-hydroxyvitamin D,” “serum 25(OH) D, OR serum vitamin D;” (2) “hip fractures OR “total fractures OR fractures;” (3) “cohort study OR case-cohort studies OR prospective study OR prospective studies” The search time is up to December 2019, and there are no restrictions on language and year of publication. To avoid initial misses, we scanned the related articles and used the “related article” function for extra searches. The full text of all citations that appeared relevant was inspected by two independent reviewers. Furthermore, we also manually searched the abstract of meetings related to Endocrinology and Traumatology, which provided printed or electronic publications. However, we did not cite these meeting abstracts in this study.All included studies were independently assessed by two reviewers. If there is any objection about data inclusion and data interpretation, it is resolved through arbitration and an agreement is reached after discussion.Characteristics of included studies are as follows: (1) observational study, (2) the patients had hip fractures or some other type of fractures and reported the corresponding serum 25(OH)D levels, or (3) calculated and reported the value of relative risks (RR), odds ratios (OR), or hazard ratios (HR) and 95% confidence interval (CIs). (4) The study population was aged ≥ 60.The exclusion criteria were as follows: (1) retrospective studies; (2) persons aged below 60 years; (3) serum 25(OH)D levels and hip or total fractures, RR, OR, or HR and 95% CI were not stated; (4) the full conference abstract was not found; and (5) the study population has a disability or other disease that affects the final outcome.Two reviewers independently extracted the data using a standardized data collection form. Discrepancies were resolved through discussion with other investigators and through reference to the original articles. The following data were extracted from each study: the first author’s last name, year of publication, type of study, country of study, gender and age of participants, years of follow-up, sample size, number of fractures, threshold of 25(OH)D levels, adjusted variables, and corresponding 95% CIs-RR estimates; if RR of different potential confounding factors is higher, the RR we extracted reflects the maximum control of potential confounding factors. If required, we contacted the authors of the preliminary study for more information.Relative risks (RR) were required as the general measure of association across studies. HRs and ORs were transformed into RRs –18.I2 statistics were used to evaluate heterogeneity between studies [I2 values of 25%, 50%, and 75% were considered low, moderate, and high, respectively [P < 0.10 values of the Cochran Q statistic, we considered statistical heterogeneity and we report a random-effects model. Subgroup analyses were conducted to assess associations between the fracture risk and relevant study characteristics  as possible sources of heterogeneity. Subgroup analysis was used for classified variables. We used funnel plot asymmetry to examine publication bias, and the Egger regression test was used to measure the asymmetry of the funnel plot [For the meta-analysis, we performed a random-effects model . Cochran studies . We follectively . For P <nel plot . We condnel plot , 24. AllThe PRISMA statement flowchart shows the process of literature screening, study selection, and reasons for exclusion Fig. . The iniIn total, 41,738 patients from 20 studies were obtained in the meta-analysis and 5916 had fractures, including 3237 hip fractures. The characteristics of the studies and participants are summarized in Table P = 0.238; I2 = 21.5%). The pooled estimate of serum 25(OH)D level and the risk of hip fracture events did not vary substantially with the exclusion of anyone study by sensitivity analysis . The pooled estimate of serum 25(OH)D level and the risk of hip fracture events did not vary substantially with the exclusion of anyone study by sensitivity analysis  shows the possibility of publication bias. We conducted the trim and fill analysis additionally , which did not vary substantially.For total fracture, Begg’s test  indicates that in the analysis by fracture site, there was no publication bias between serum vitamin D levels and hip fracture events. The trim and fill analysis show the similar results D levels and the risk of fracture was conducted. The results are shown in Table This meta-analysis was based on a total of 41,738 participants and 5916 patients (including 3237 hip fractures) who were older than 60 years indicated that serum 25 hydroxyvitamin D levels were negatively correlated to the risk of hip fracture. However, serum 25(OH)D has no association with total fracture risk. To our knowledge, this is the first meta-analysis of the relationship between serum vitamin D levels and fracture risk using age as an inclusion criterion. We researched the literature and browsed all relevant literature on total and hip fractures and we excluded all studies that did not meet the age requirements.A few meta-analyses have been performed in this field, but they showed different results. Lv et al.  concludeHowever, it was noteworthy that because of the different bone turnover rates, osteoporotic fractures in the elderly are divided into perimenopausal osteoporotic fractures and senile osteoporotic fractures. Evaluating these two types together adds uncertainty to the conclusion. Compared to the two meta-analyses above, our study focused on the population of elderly persons aged ≥ 60 years. These two factors may be the reasons for different conclusions.Efficacy of vitamin D supplements in the elderly to prevent osteoporotic fractures is still uncertain. Heike et al.  conducteIn recent years, serum markers and fracture risk have become a hot field with interest increasingly. Osteoporosis (OP) is a common and silent disease, and it is the primary cause of pathological fractures in the elderly . The linThe inverse relationship between low serum 25(OH)D and fractures might have several plausible mechanisms. First, severe vitamin D deficiency causes rickets or osteomalacia, the consequences of vitamin D deficiency can cause osteoporosis and fractures, mineralization defects, which may lead to osteomalacia in the long term, and with muscle strength decreasing, falls and fractures will eventually occur. Second, vitamin D deficiency also causes higher secretion of parathyroid hormone (PTH) which leads to high bone turnover and increased bone resorption and eventually it will lead to bone loss or fracture. Hence, on the one hand, severe vitamin D deficiency causes a mineralization problem and osteomalacia and on the other hand increased PTH content can lead to increased bone conversion, bone resorption, and osteoporosis of bone. Third, several studies have shown a positive correlation between serum vitamin D and BMD –45; low The strengths of our meta-analysis are as follows: First, our study included a total of 41,738 participants and 5916 fracture events (including 3237 hip fractures), which greatly improved the statistical power of the analysis. Second, our evaluation criteria are based on observational studies. On the one hand, recall and selection bias in case-control studies can be reduced; on the other hand, it also provides additional data on the risk of fracture among participants. Third, all included studies were independently assessed by two reviewers. In case of discrepancies or disagreements pertaining to research inclusion and data interpretation, we resolved them through arbitration and discussion to reach a final agreement. Therefore, the errors in data were reduced. Fourth, all studies had a long period of follow-up and high scores in literature quality assessment .However, the limitations of our study must be considered. First, despite the RR adjustment and the high-quality assessment scores of studies, our study is still influenced by a number of confounding factors that could be inherent in the obtained cohorts, which is a mutual disadvantage of all observational studies and meta-analyses, which can cause deviations in risk estimates. The serum vitamin D level is related to the sunlight duration. Due to the lack of data and the characteristics of prospective experiments, it is difficult to estimate the normal exposure time of participants, which may affect the final serum vitamin D test results. Second, differences in research methodology may be a source of heterogeneity. In the subgroup analysis of this study, we founded gender, location, the year that the study started, and analysis the sources of heterogeneity through these subgroups. Despite these factors that may reduce the strength of the conclusion, our forest plots indicate that relative risk is fairly consistent throughout the study.Our study indicates that compared to low serum 25(OH)D levels, high serum vitamin D protects against the risk of hip fracture in patients ≥ 60 years old with osteoporosis. However, serum 25(OH)D levels were not correlated to the total fracture risk. Although serum vitamin D levels cannot directly affect the risk of fractures, the indirect effects of low serum vitamin D levels on fractures suggest that vitamin D supplementation is still necessary.


Evidence on the carcinogenicity of coalmine dust in occupational settings is still conflicting. Therefore, we conducted this research to evaluate the mortality risk of lung cancer for coalminers exposed to occupational dust when compared to population with no or low dust exposure.  Databases of PubMed and Chinese National Knowledge Infrastructure as well as reference lists were searched updated to September 18, 2020. The enrolled articles should report lung cancer mortality risk for coalminers exposed to occupational dust. Basic information was extracted such as the author and publication year, area and ethnicity, the type and estimates of outcome, duration of follow-up, and the study design. The checklists from Agency for Healthcare Research and Quality and the Newcastle-Ottawa Scale were used for the assessment of quality and bias risk for descriptive studies, cohort studies, and case control studies, respectively. The overall relative risks were calculated while Begg's and Egger's tests and sensitivity analysis were performed to explore potential heterogeneity sources. Metaregression and subgroup analyses were also conducted to give more detailed information for the correlation between dust exposure and lung cancer mortality. I2 = 95%, P < 0.001). The pooled relative risk of mortality from lung cancer was 1.16 (1.03-1.30) for coalminers. Results of metaregression analysis indicated that the high heterogeneity among these enrolled studies might be caused by the ethnicity differences (P = 0.011). Subgroup analysis also indicated that the pooled estimate for Asian population in China was 4.94 (3.95-6.17) with I2 = 39.3% and P = 0.192. All these results suggested that exposure to occupational dust would significantly increase the mortality risk of lung cancer, especially for Asian population in China, which should be measured and controlled more strictly. Discussion. This systematic review and meta-analysis provides high-quality evidence that exposure to occupational dust might increase the mortality risk of lung cancer, especially for Asian populations in China. The magnitude of this effect is of major public health importance in view of the ubiquitous existence of coalmining industry in China and even in the world. However, these pooled estimates should be interpreted cautiously because of the high heterogeneity among these publications. Other. This study was supported by the National Key Research and Development Program of China (Grant No. 2016YFC1302501). A total of 19 articles with 22 different studies  including 8909 observed deaths from 1964 to 2017 were enrolled with a significant heterogeneity ( There is an increasing urgency in estimating the burdens of different diseases, especially of different cancers , 2, whicCoalminers tend to be exposed to a series of risk factors during work such as mine dust, vibrations transmitted to the hands, noise, and a high heat load as well as an overloading of the upper extremities when working with pneumatic tools. Among these risk factors, mine dust containing crystalline forms of silica might be the most serious one that can significantly affect the health and life of coalminers.In recent years, death from respiratory diseases is still an important occupational hazard for coalminers. The relationships between occupational exposures to coalmine dust and mortality from coal workers' pneumoconiosis and chronic obstructive pulmonary disease have been clearly established –5. The mTherefore, this current work was conducted to evaluate the mortality risk of lung cancer for coalminers exposed to occupational dust when compared to population with no or low dust exposure, which can facilitate the development of coalminers' risk reduction and health management. In addition, we also aimed to explore the differences of mortality risk of lung cancer for coalminers from different ethnicities, study designs, and outcome types based on the subgroup analyses.This analysis was reported based on the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement and checklist. Both PubMed and Chinese National Knowledge Infrastructure (CNKI) databases were searched for relevant publications. The full search strategy without any limits for these two databases was as follows:  and (lung tumor or lung carcinoma or lung neoplasm or lung cancer) and dust, updated to September 18, 2020. Reference lists of the enrolled publications were also retrieved for acceptable literatures.A series of criteria were drawn up for the inclusion and exclusion of publications, which were in accordance with the goal of this study. For the literature inclusion, the criteria were as follows: (1) target populations enrolled in the studies were coalminers exposed to occupational dust while the compared participants were individuals with no or low dust exposure; (2) lung cancer mortality should be reported as an outcome; (3) the original estimates of mortality along with 95% confidence intervals (95% CI) in the form of standardized mortality ratio (SMR), proportional mortality ratio (PMR), odds ratio (OR), or relative risk (RR) were reported in the articles; (4) the enrolled studies should be descriptive studies, case control studies, or cohort studies. For the exclusion, the criteria were as follows: (1) unrelated publications, reviews, case reports, and comments; (2) articles shared the same cohort or population; and (3) estimates of outcomes and the 95% confidence intervals as well as other information that are needed by the synthesized meta-analysis were missing. For studies including a wider population than our inclusion criteria, the eligible participants would be enrolled in our analysis while ineligible participants would be excluded. Overall, the process for determining which studies were eligible for inclusion could be concluded as publication search, exclusion for reviews, case reports, comments, unrelated and duplicated articles as well as articles without complete information through the full text scanning, and then the final remaining publications would be enrolled in this meta-analysis.The information that was needed during the meta-analysis at both study level and participant level was extracted from the included publications by two authors independently as listed: the first author, publication year, participants' age, gender, area and ethnicity, number of observed deaths, the type and estimates of outcome with 95% CI, duration of follow-up, type of control and case groups, and the study design. Moreover, the duration of follow-up from different publications was unified by year, while the estimates of outcome were standardized in the form of mortality ratio.The methodological quality of the descriptive studies was assessed using the 11-item checklist recommended by the Agency for Healthcare Research and Quality (AHRQ), which included the definition of information source, inclusion and exclusion criteria, time period and continuity for identifying patients, blinding of personnel, assessments for quality assurance, confounding and missing data, and patient response rates and completeness, while the Newcastle-Ottawa Scale (NOS) was followed for the assessment of quality and bias risk for cohort studies and case control studies. For case control study, the items referred to the definition, selection and representativeness of cases and controls, comparability of cases and controls, ascertainment of exposure, and nonresponse rate, while for cohort study, selection and representativeness of the cohorts, ascertainment of exposure, assessment of outcome during the study, comparability of cohorts, and adequacy of follow-up were the major assessment items. An item would be scored “0” if it was answered “UNCLEAR” or “NO”; for the answer of “YES”, the item would get a “1” score. Article quality for descriptive study was assessed as follows: low quality: 0-3; moderate quality: 4-7; and high quality: 8-11. The score on the NOS ranged from 0 to 9 with a score equal to or greater than 6 identified as a feasible methodological design. For study with low quality or significant risk of bias, it would be eliminated from the following data synthesis.Q test and I2 value [P value for the Q test less than 0.05 or I2 > 50% indicated that there was significant heterogeneity, and then, the overall estimate with 95% CI was pooled with the random-effect models. Otherwise, we would conduct the data synthesis with the fixed-effect models. In order to avoid the bias in the process of data synthesis caused by the fact that studies with large samples and positive results were easier to be published and retrieved than studies with small samples or negative results, Begg's and Egger's tests were performed to assess the existence of publication bias in this study, which would not be considered existed when the points representing the literature were distributed symmetrically or uniformly on both sides of the midline and mainly concentrated at the funnel tip and P > 0.05 for the Egger test. In addition, the sensitivity analysis was carried out by excluding the literature one by one. If the pooled estimates and 95% confidence intervals were evenly distributed on both sides of the midline and closely arranged, we can conclude that the research was stable and reliable; otherwise, the literatures deviating from the midline seriously should be excluded to improve the quality of the meta-analysis. Finally, subgroup analyses and metaregression analysis were conducted to identify potential sources of the heterogeneity, which were grouped by type of outcome: SMR, PMR, OR, and RR; ethnicity: Asian, Caucasian, and Caucasian/Negro; control group: general population, coalminers, and quarrymen; case group: coalminers or coalminers with pneumoconiosis; type of study design: cohort study, case control study, and descriptive study; number of observed deaths: <100 and >100; year of publication: 1964-1980, 1980-2000, and 2000-2017; duration of follow-up: <10 years, >10 years and <20 years, and >20 years; area: China, UK, US, and others; and study quality: moderate quality and high quality. All statistical analyses were performed by STATA 11.0 , and results of P < 0.05 were considered as statistically significant.Lung cancer mortality risk was prespecified as the primary outcome in this study, which was compared between coalminers exposed to occupational dust and general population without dust exposure or occupational population with low dust exposure in the form of SMR, PMR, OR, or RR with their 95% CIs. This study used a two-stage approach that first generated estimates of effectiveness for each study separately and then combined these summary statistics, which can be shown from the forest plots. The heterogeneity among these enrolled studies was assessed by I2 value . P valueAs shown in I2 = 95%, P < 0.001). The pooled mortality estimate was 1.16 (95% CI: 1.03-1.30) with P = 0.013 as shown in The random-effect model was performed since significant heterogeneity existed among these 22 studies . In the subgroup of the ethnicity, the pooled estimate for the Asian population was 4.94 (3.95-6.17) with P < 0.001; however, exposure to dust might not increase the mortality risk of lung cancer in the Caucasian and Negro population. When compared with the general population, coalminers exposed to occupational dust would be susceptible to the mortality of lung cancer with a risk estimate of 1.23 (1.08-1.40). In addition, coalminers who worked during 1980 and 2000 obtained a higher risk  compared to the coalminers before 1980 and after 2000. In the cohort studies, coalminers were more susceptible to death from lung cancer with a relative risk of 1.42 (1.10-1.85). Studies from China indicated a higher risk for coalminers with RR of 4.94 (3.95-6.17). For miners with duration of follow-up less than 10 years, the mortality risk was 1.26 (1.00-1.56). In addition, there was a statistical significant correlation between dust exposure and lung cancer mortality with the relative risk of 1.28 (1.07-1.53) in the studies with high quality. But such a correlation was not found in the studies with moderate quality. With respect to the other types of subgroup analyses including the case and control groups, types of the study design, and observed deaths, there were no statistical significant correlations between exposure to occupational dust and mortality risk of lung cancer.A multivariate metaregression was conducted to explore the potential factors of the heterogeneity in the pooled estimate, and results are summarized in Overall, these results suggested that exposure to occupational dust would significantly increase the mortality risk of lung cancer, especially for Asian population in China, which prompted that relevant departments should take more strict measures to measure and control occupational dust and strengthen the occupational protection and health promotion for coalminers.Although a global movement toward alternative energy has been taken, coal still undertakes nearly a third of worldwide energy supply , and thiI2 = 95%, P < 0.001). Therefore, we conducted the sensitivity analysis and metaregression analysis to find the potential heterogeneity sources, and results indicated that the high heterogeneity might result from the different ethnicities. Different risks were also observed among different ethnicities in the following subgroup analysis. The Asian populations in China tended to be susceptible to lung cancer mortality with a pooled risk estimate of 4.94 (3.95-6.17), while this association was not found for Caucasian or Negro populations. This risk difference might result from different data sources, statistical methods, and working conditions. Since publications for the Asian populations were conducted in China from 1988 to 1994, which was a period with rapid economic development due to the implementation and progress of reform and opening up. Great progress and rapid development have been made in industrial production, especially in the coal mining industry. However, the working conditions, monitoring, and control of pollutants as well as the awareness of own protection for coalminers have not been improved accordingly. Therefore, the morbidity and mortality risk of various respiratory diseases rose considerably among coalminers. Based on subgroup analysis for the year of publication, we observed an excessive mortality risk of 1.56 (1.10-2.20) from 1980 to 2000. In addition, there was a statistical significant correlation between dust exposure and lung cancer mortality with the relative risk of 1.42 (1.10-1.85) in cohort studies, especially in studies with high quality . Result from the subgroup analysis for Asian populations from China can be strongly convincing since no significant heterogeneity existed . However, there was high heterogeneity in the data synthesis for both overall and all other subgroup analysis with I2 > 50% and P < 0.05, so these results should be interpreted with caution by the relevant groups and researches.Overall, this present meta-analysis, which combined the results from 22 different studies, supports the carcinogenicity of coalmine dust on the lung with the pooled risk estimate of 1.16 (95% CI: 1.03-1.30). However, there was a significant heterogeneity among these enrolled researches (Strengths of our meta-analysis included its comprehensive and up-to-date search for both the electronic databases and reference lists of the enrolled publications. In addition, more than 8909 events from varied areas, ethnicities, and study designs were enrolled in this analysis, which can decrease the risk of bias overall to provide a more convincing and generic conclusion.Despite the obvious strengths, some limitations still existed in this study. Firstly, limitations of this study included the variation in study settings and individual participants, which might cause the high heterogeneity among these publications. In addition, publications without complete information at individual participant level or study level were removed in our meta-analysis, resulting in a limited number of enrolled studies and sample size. Although metaregression and subgroup analyses were performed, the pooled estimates should be interpreted cautiously since there is high heterogeneity among these publications. Therefore, prospective large cohort studies are warranted to verify the results of this study that were attained through pooled analysis of 22 studies conducted in varied protocols and settings.To conclude, this systematic review and meta-analysis supports the positive association between coalmine dust and mortality risk of lung cancer, especially for Asian populations from China when compared to the general population during 1980 and 2000. The magnitude of this effect is of major global public health importance in view of the ubiquitous existence of the coalmining industry worldwide, which should prompt relevant departments to take more strict measures for the measurement and control of the occupational dust and the protection for coalminers.


PubMed, Web of Science, and the Cochrane Database of Systematic Reviews were searched for meta-analyses that provided risk estimates  for associations between intakes of whole and refined grains and risk of total and site-specific cancer. The preferred reporting items for systematic reviews and meta-analyses (PRISMA) guidelines were followed. Only meta-analyses that included whole grains and refined grains as separate food groups, and not as part of dietary patterns, were included. A total of 17 publications were identified that met inclusion criteria. Within these, results from a total of 54 distinct meta-analyses were reported for whole grains and 5 meta-analyses for refined grains. For total cancer mortality, 7 meta-analyses of cohort studies indicated that whole grain intake was associated with 6% to 12% lower risk in comparison of highest vs. lowest intake groups, and 3% to 20% lower risk for doses ranging from 15 to 90 g/day. For site-specific cancers, meta-analyses indicated that whole grain intake was consistently associated with lower risks of colorectal, colon, gastric, pancreatic, and esophageal cancers. Limited data were available for refined grains, with only 4 publications providing risk estimates, and only 1 of the meta-analyses included more than 3 studies. High intake of refined grains was associated with increased risk of colon and gastric cancer. By contrast, in the only dose-response meta-analysis, each 90 g/day consumption of refined grains was associated with a 6% lower risk of total cancer. In addition to the limited number of published meta-analyses on refined grains, results were also weakened due to the fact that refined grains were frequently defined to include both staple grain foods and indulgent grain foods, and the majority of studies included in the meta-analyses provided no specific definition of refined grains. Overall, meta-analyses of cohort and case-control studies consistently demonstrate that whole grain intake is associated with lower risk of total and site-specific cancer, and support current dietary recommendations to increase whole grain consumption. By contrast, the relationship between refined grain intake and cancer risk is inconclusive. Whole grains are associated with reduced risk of a number of chronic diseases, including cancer, and are recommended as an important part of a healthy diet ,2,3. By Much of the published research used to inform dietary guidelines comes from cohort studies that have examined the association between dietary patterns and risk of various chronic diseases . A healtTherefore, the objective of this review was to summarize the published research on the association between whole grain intake and cancer risk and between refined grain intake and cancer risk . The review focuses exclusively on results from meta-analyses of observational cohort and case-control studies of adults in which whole grains and refined grains were evaluated as distinct food categories and not as part of dietary patterns.The preferred reporting items for systematic reviews and meta-analyses (PRISMA) guidelines were used for this review . The InsThe flow chart for publication selection is presented in Details of the meta-analyses, including information on the number of cohort and case-control studies included, total number of participants, and results of publication bias and heterogeneity assessments, are presented in 2 statistic [2 between 53% and 91%  . Cohortstatistic . Twenty- and 91% .Eight meta-analyses have been published on the relationship between whole grain intake and either cancer mortality ,29,31,33In the one meta-analysis that relied solely on case-control studies , a pooleIn summary, even though 8 meta-analyses have been published on the association between whole grain intake and total cancer mortality, it must be noted that 6 of meta-analyses ,26,29,33Eleven of the published meta-analyses provided risk estimates for whole grain intake and site-specific cancer risk, including colorectal ,26,28,32Without exception, all six meta-analyses reported significantly lower colorectal cancer risk associated with whole grain intake . Four ofFor colon cancer, all three meta-analyses reported lower risk associated with whole grain intake. In the highest vs. lowest intake comparisons, high intake of whole grains was associated with a 15% to 18% lower risk of colon cancer. In the dose-response analyses, colon cancer risk was reduced by 14% to 18% for each 90 g/day ,28 and bBy contrast, for rectal cancer only one of the three meta-analyses reported a significantly lower risk associated with whole grain intake. Although the relative risks in categorical and dose-response analyses ranged between 0.80 and 0.94, only the 20% lower risk reported by Schwingshackl et al.  in the hFour meta-analyses have been published on the association between whole grain intake and gastric cancer risk ,32,35,36Two meta-analyses have been published on the association between whole grain intake and pancreatic cancer risk. In a meta-analysis of 4 case-control studies, Jacobs et al.  reportedThree meta-analyses have been published on the association between whole grain intake and risk of prostate cancer, including 2 categorical analyses ,34 and 1Three meta-analyses have been published on breast cancer, including 2 categorical analyses ,30 and 1In an earlier meta-analysis by Jacobs et al. , which iBoth meta-analyses on esophageal cancer, using mostly  or entirLimited data have been published on other cancers, including oral, brain, endometrial, and non-Hodgkin’s lymphoma. For each of these cancers, only one meta-analysis has been published, using case-control studies . In cateIn contrast to whole grains, only a few meta-analyses have been published on the association between refined grain intake and cancer, including total cancer , colorecIn a dose-response meta-analysis by Aune et al. , which iSchwingshackl et al.  includedTwo meta-analyses have reported on the association between refined grain intake and risk of gastric cancer. Using highest vs. lowest intake comparisons from 18 case-control studies, Wang et al.  reportedPublished meta-analyses reviewed herein demonstrate that whole grain intake is consistently associated with lower risk of total cancer mortality. This was observed in both categorical and dose-response analyses. Consistently lower risks for colorectal, colon, gastric, pancreatic, and esophageal cancers were also observed. These findings are congruent with the findings of expert reports and public health recommendations that advocate for greater consumption of whole grains ,3,10. MaIn contrast to these findings, whole grain intake was not associated with decreased risk of prostate cancer. Only two publications reported meta-analyses on whole grain intake and prostate cancer risk . The onlUnlike the results for whole grains, meta-analyses revealed no consistent findings on the association between refined grain intake and cancer risk. Three of the meta-analyses indicated that refined grain intake may be associated with higher risk of colon and gastric cancer. In contrast, one meta-analysis reported a 6% lower risk of total cancer mortality associated with higher intake of refined grains. There are several important limitations to the meta-analyses on refined grains, and these are discussed below. Nevertheless, the limited and inconsistent findings from these meta-analyses on refined grains analyzed as a separate food category are not entirely supportive of the data from dietary pattern analyses showing that a Western dietary pattern that includes refined grains is usually associated with higher risk of cancer ,13,14,15Whole grains may reduce cancer risk via a number of mechanisms, as reviewed previously ,157,158.Not all meta-analyses and systematic reviews show uniformly lower cancer risks associated with cereal fiber intake ,171,172.A major weakness in the literature summarized in this review is the lack of consistency in the definitions of whole grains and refined grains in studies included in the meta-analyses . AlthougWhereas whole grains are typically defined to include staple grain foods such as bread, cereals, and pasta, many studies have included indulgent grain foods in their definition of refined grains . In the On the other hand, refined grain foods are an important source of dietary fiber due to the fact that refined grain consumption is much greater than whole grain consumption ,176. BasIt is also important to acknowledge that 6 of the 9 meta-analyses on total cancer mortality relied on essentially the same 6 cohort studies for determination of relative risks ,84,85,872, which suggests that variation across studies was not due to chance. The considerable differences in definitions of whole and refined grains may also have contributed to the significant heterogeneity across studies. Despite these weaknesses, the meta-analyses reviewed have some notable strengths. With two exceptions, all meta-analyses reported no publication bias that might confound interpretation of the results. The only meta-analysis that reported a significant Egger’s test was that of Zhang et al.  on esophMeta-analyses consistently show that whole grain consumption is associated with lower risk of total cancer mortality. Risk reductions for the highest intakes of whole grains ranged between 5% and 12%. In dose-response analyses, each 30 g/day intake of whole grains was associated with a ~7% lower risk of cancer mortality. For site-specific cancers, meta-analyses indicate that whole grain intake is consistently associated with lower cancer risk, with the strongest evidence for colorectal, gastric, pancreatic, and esophageal cancers. The only cancer for which whole grain intake was not associated with lower risk was prostate cancer. Overall, these meta-analyses of cohort and case-control studies support the recommendations for increased whole grain consumption [Refined grain intake may reduce the risk of total cancer, but the meta-analysis included only 2 studies. High intake of refined grain may increase risk of gastric cancer, but these conclusions must be viewed skeptically due to the weaknesses described above. Considerably more research is necessary on the association between refined grain intake and cancer risk before definitive conclusions can be drawn and evidence-based dietary guidelines established.


Therefore, we performed this meta-analysis to further investigate whether TNF-α variants are related to AKI susceptibility.  Several studies have reported conflicting findings regarding the association between tumor necrosis factor-alpha  and 95% corresponding confidence intervals (95% CIs) were analyzed to evaluate the strength of the relationship.  A comprehensive search of observational studies on the association of TNF-α rs1800629 variant and AKI (A vs. G: OR [95%CI] = 1.33 [0.98‐1.81]) among the overall population. Ethnicity-stratified analysis indicated that the TNF-α rs1800629 variant was a risk factor for Asians (OR [95%CI] = 1.93 [1.59‐2.35]) while it is not for Caucasians (OR [95%CI] = 1.04 [0.91‐1.20]). Additionally, we also found that TNF-α rs1799964 polymorphism was observed to have a significant relationship with AKI risk in Asian patients .  A total of 8 studies involving 6694 patients (2559 cases and 4135 controls) were included. Pooled analysis showed a trend of increased risk between the TNF-α rs1799964 polymorphism was also related to a significantly increased risk of AKI in Asians. The TNF rs1800629 polymorphism exhibited a trend toward AKI susceptibility with ethnic differences. The relationship was found to be significant among the Asian population, but not among those of Caucasian origin. Additionally, the TNF- Acute kidney injury (AKI), characterized by a rapid decline in kidney function within two days, has become one of the most common syndromes in hospital settings worldwide . AKI occα (TNF-α) could initiate the inflammatory cascade and induces the production of inflammatory mediators  = 1.33 [0.98‐1.81]), an increasing trend of AKI risk could be seen, as shown in b) = 1.85; P = 0.004; adjusted R2 = 100%, see In order to explore the possible sources of heterogeneity, meta-regression analysis was performed. The result indicated that ethnicity had a significant association with outcomes at posttest . Symmetrical funnel plots are shown in α rs1799964.Publication bias was evaluated by Begg's funnel plot and Egger's regression test. The results showed that no significant publication bias was found in the TNF-α rs1800629 was significantly associated with AKI susceptibility in Asians and but not in Caucasians. We next investigated the allele frequencies of TNF-α rs1800629 between ethnicity  was different from that in the 1000 Genomes Project (0.196 for Asians).As shown above, TNF-city see . In our α polymorphisms and AKI susceptibility. The results indicated that TNF-α rs1800629 had a trend of increased risk of AKI for all patients. What is more, this association was significant in Asian patients, but was not significant in Caucasian patients by subgroup analysis. Additionally, because of the limited number of included studies, TNF-α rs1799964 was shown to be related to an increased risk for AKI only in Asian patients.To our knowledge, the present study is the first global systematic meta-analysis to evaluate the genetic relationship between TNF-α polymorphisms and AKI susceptibility was controversial. In comparison, both of them did not assess the effect of TNF-α polymorphisms or conduct a detailed analysis of the conflicting results. Consequently, we performed this meta-analysis to quantificationally evaluate the relationship between TNF-α polymorphisms and AKI susceptibility. More importantly, we found that TNF-α polymorphisms had a trend of increasing the risk of AKI susceptibility with ethnic differences. In the present meta-analysis, the association between TNF-α rs1800629 and AKI susceptibility was not significant for the overall populations. Meta-regression and ethnicity-stratified analyses were performed to investigate the main source of heterogeneity. The results showed that the association of TNF-α rs1800629 with AKI susceptibility varied by ethnicity. What is more, the subgroup results indicated that the rs1800629 GA/AA genotype had a higher risk of AKI among Asians but not among Caucasians. Several possible explanations may account for these findings. One of the reasons is that the studies conducted in Caucasians mainly included critically ill patients from the intensive care unit with severe infection [α rs1800629 alleles is population dependent. In the present meta-analysis, the frequencies of the rs1800629 A allele in the control group were 6.26% for Asians [In the previous systematic reviews, both Vilander et al.  and Laranfection , 19–21 onfection . The patnfection  or were nfection , with a r Asians , 17 and r Asians , 19–21. r Asians . Moreover Asians , dilatedr Asians , and chrα rs1799964, a total of 3 eligible studies conducted in the Asian population including 1204 cases and 1881 controls were analyzed. Although there was a difference of the TNF-α rs1799964 allele frequency observed between our meta-analysis and the 1000 Genomes Project data, there was no deviation from HWE in the control group of all included studies. The difference may be due to the small sample size or the heterogeneity between the included studies. The relationship between TNF-α rs1799964 and AKI susceptibility was found in all the genetic comparisons, suggesting that the rs1799964 variant might lead to a higher risk for AKI in the Asian population, while the relationship needs to be verified in a larger sample size.Regarding TNF-α polymorphisms on AKI susceptibility by screening in the online database, while it is difficult for us to get access to the raw data of AKI genomics to make the validation. Second, AKI is a multifactorial disease. Some confounding factors such as gender, history of nephrotoxic drugs, and heart function may be addressed across the included studies. However, we have no access to get these clinical data to estimate the effect of confounding factors. Third, significant heterogeneity was found in the rs1800629 analysis for the overall population. Thus, we performed meta-regression and ethnicity subgroup analyses to investigate the source of the heterogeneity. Last, the included studies for TNF-α rs1800629 subgroup analysis or rs1799964 analysis were relatively limited; the results of these analyses should be interpreted prudently.Nevertheless, some limitations of this study should not be ignored. First, it is better to further evaluate the effect of TNF-α rs1799964 polymorphism was related to a significantly increased risk of AKI in Asians. However, as AKI is a multifactorial disease, our findings require verification in larger samples and functional studies in the future.In conclusion, TNF rs1800629 polymorphism had a trend of increasing the risk of AKI with ethnic differences. The trend was significant among Asian patients, but was not among Caucasian patients. Additionally, the TNF-