Many researchers have shown that pretreatment plasma fibrinogen levels are closely correlated with the prognosis of patients with lung cancer (LC). In this study, we thus performed a meta-analysis to systematically assess the prognostic value of pretreatment plasma fibrinogen levels in LC patients.A computerized systematic search in PubMed, EMBASE, Web of Science and China National Knowledge Infrastructure (CNKI) was performed up to March 15, 2018. Studies with available data on the prognostic value of plasma fibrinogen in LC patients were eligible for inclusion. The pooled hazard ratios (HRs) and odd ratios (ORs) with 95% confidence intervals (CIs) were used to evaluate the correlation between pretreatment plasma fibrinogen levels and prognosis as well as clinicopathological characteristics.p=0.001), disease-free survival (DFS) , and progression-free survival (PFS) . Furthermore, our subgroup and sensitivity analyses demonstrated that the pooled HR for OS was robust and reliable. In addition, we also found that a higher fibrinogen level predicted advanced TNM stage (III-IV)  and a higher incidence of lymph node metastasis .A total of 17 studies with 6,460 LC patients were included in this meta-analysis. A higher pretreatment plasma fibrinogen level was significantly associated with worse overall survival (OS) (HR: 1.57; 95% CI: 1.39-1.77; Our study suggested that higher pretreatment plasma fibrinogen levels predict worse prognoses in LC patients. Lung cancer (LC) remains a major cancer-associated cause of death worldwide. LC falls into either non-small cell lung cancer (NSCLC) or small cell lung cancer (SCLC), with NSCLC accounting for nearly 85% of incidents ,2. AlthoAn increasing amount of evidence has shown that there is a close association between the activation of coagulation and fibrinolysis and cancer invasion, angiogenesis and metastasis -13. AddiWe systemically searched the accessible databases, including PubMed, Web of Science, EMBASE and China National Knowledge Infrastructure (CNKI), to identify eligible studies published up to March 15, 2018. Three key search items were used during this process: (fibrinogen or hyperfibrinogenemia), (lung cancer/carcinoma or pulmonary cancer/carcinoma) and . The search strategy used in PubMed was as follows: search ((fibrinogen[Title/Abstract]) OR hyperfibrinogenemia[Title/Abstract]) AND ((((lung cancer[Title/Abstract]) OR lung carcinoma[Title/Abstract]) OR pulmonary cancer[Title/Abstract]) OR pulmonary carcinoma [Title/Abstract]) AND ((((prognosis[Title/Abstract]) OR prognostic[Title/Abstract]) OR survival[Title/Abstract]) OR outcome[Title/Abstract]). Additionally, we manually checked the reference lists in relevant reviews and retrieved studies to identify additional potentially eligible studies.The inclusion criteria were as follows:  clinicalTwo authors extracted the relevant data in an independent manner. The extracted data included the first author, publication year, country, recruitment period, median age, study sample size, histological type of LC, tumor TNM stage, cut-off values for high fibrinogen level, treatment type, follow-up time, HRs with 95% CIs for survival outcomes, and analysis type. During this process, any discrepancies were resolved by discussion among all the authors.In this meta-analysis, we applied the Newcastle–Ottawa Scale (NOS) score to evaluate the quality of each included study . The NOSp-value <0.05 indicated worse OS and more unfavorable clinicopathological parameters. We assessed the statistical heterogeneity across the studies using Cochran’s Q test and Higgins I2 statistic. A random-effects model was used to calculate the parameters when there was significant heterogeneity across studies (I2>50% indicated high heterogeneity). Otherwise, a fixed-effects model was applied. Subgroup analysis, stratified by country, sample size, TNM stage, cut-off value, treatment type, starting time of recruitment period and analysis type, was performed to investigate the potential sources of heterogeneity and to concurrently test the reliability of the total pooled results. In addition, we performed sensitivity analysis to further test the robustness of the total pooled results by the sequential omission of individual studies. Publication bias was evaluated using Begg’s funnel plots  and their corresponding 95% CIs were used to quantitatively determine the link between plasma fibrinogen and the clinicopathological parameters of patients with LC. The coexistence of HR or OR>1 (low fibrinogen as reference), 95% CI not overlapping 1, and el plots  and Eggeel plots . If one The study identification process is summarized in Our meta-analysis included a total of 6,460 patients from China n=15) ,25,26,31 ,25,26,3p=0.001; p=0.003; p<0.001; Of the included studies, 16 studies reported OS -26,30,31p<0.001; p=0.02; p=0.08; p=0.11; In this meta-analysis, sufficient data were available for analyzing the association between the fibrinogen level and several clinicopathological parameters, including sex -18,20,24To investigate the possible sources of heterogeneity for the pooled HR of OS, subgroup analysis was performed according to country, sample size, recruitment time, cut-off value, treatment type, histological type and analysis type. From the results, we observed that significant heterogeneity did not continuously exist in the subgroups of sample size, recruitment period and treatment type , suggestp-value <0.05 was obtained in Egger’s test, indicating that significant publication bias existed among these studies exploring the association between the fibrinogen level and OS in LC. To evaluate the degree of the influence that the publication bias had on the dependability of the pooled HR for OS, the trim-and-fill method was used in this analysis. The results showed that the adjusted funnel plot became symmetric (Begg’s test and Egger’s test were used to evaluate publication bias for OS. Our results showed that Begg’s funnel plot appeared asymmetric , and a pymmetric . More imThe association between pretreatment plasma fibrinogen levels and prognosis has been investigated in a variety of malignant tumors, including LC. However, no meta-analyses have been conducted to systematically assess the correlation between the plasma fibrinogen level and prognosis as well as the clinicopathological characteristics of LC patients. To the best of our knowledge, this study is the first meta-analysis to systematically evaluate the prognostic value of pretreatment plasma fibrinogen levels in LC.In this meta-analysis, a total of 17 eligible studies were finally included, of which 14 focused on NSCLC, 2 focused on SCLC and 1 focused on the mixed populations of NSCLC and SCLC. We found that a higher fibrinogen level was significantly associated with worse OS, DFS and PFS in LC patients. Furthermore, our subgroup analysis and sensitivity analysis verified the robustness and reliability of our pooled HR for OS. In particular, the multivariate analysis of the subgroups showed that the pooled HR for OS continued to be over 1, with its CI not overlapping 1, which indicated that the pretreatment plasma fibrinogen level may be an independent prognostic factor for LC patients. In addition, we also found that a higher fibrinogen level was significantly related to advanced TNM stage and a higher frequency of lymph node metastasis.Several biological mechanisms may explain the prognostic significance of higher pretreatment plasma fibrinogen levels in LC patients. First, fibrinogen is a biomarker that could reflect the status of the tumor-associated inflammatory response . Many tuApart from the interesting findings, our analysis has several limitations as well. First, studies published in only English and Chinese were included, and studies published in other languages were not considered, which may introduce publication bias. Second, the estimated HR was abstracted from the KM survival curve due to a lack of original data, likely causing some statistical errors and affecting the precision of the estimation. Third, the included studies had inconsistent TNM stages and different accrual periods, therefore leading to some heterogeneity in our meta-analysis. Fourth, the majority of the included studies were conducted in China, so more clinical studies should be performed to assess the association between plasma fibrinogen and the prognosis of LC patients from other regions. There may be three potential reasons why most of the included studies were performed in China. First, LC is the leading cause of cancer-related death, and its incidence and mortality in China have substantially increased in recent years as a consequence of lifestyle diversity and socioeconomic development . In partIn conclusion, our study showed that high pretreatment plasma fibrinogen levels predicted a worse prognosis in LC. However, more prospectively well-designed studies are needed to further confirm this conclusion due to some aforementioned limitations.Huang Z and Zhang Y conceived and designed the study. Lin G and Huang Y performed the literature search. Li R and Cao J performed the statistical analysis. Zhang Y and Huang Y wrote the manuscript. Dong M and Huang Z reviewed and edited the manuscript. Deng Y revised the manuscript. All authors read and approved the final version of the manuscript.


Increasing evidences indicated that diabetes might increase the incidence of gallbladder cancer. However, no sufficient data has ever clarified the impact of diabetes on the survival of patients with gallbladder cancer.We comprehensively searched PubMed, Embase, and the Cochrane Library databases through July 2019 in order to find sufficient eligible researches. The pooled hazard risks (HRs) and relative risks (RRs) with 95% confidence intervals (CIs) were calculated with either fix-effects or random-effects model. Due to the low gallbladder cancer mortality in general population, the RRs and standard mortality ratios (SMRs) were considered the similar estimates of the HRs.P < 0.00001). However, the mortality in male diabetes patients was not higher than female patients .Ten eligible studies were included in this meta-analysis. Analysis of eight cohorts found that diabetes was closely associated with the mortality of gallbladder cancer (HR = 1.10; 95% CI: 1.06–1.14; These findings indicated that diabetes patients had a higher mortality of gallbladder cancer compared with non-diabetes. Gallbladder cancer (GBC) is one of the most common biliary tract malignancies worldwide . By and Diabetes mellitus (DM) is a costly chronic disease worldwide. The incremental increase in costs of this disease have laid economic burdens on both financial expenditure in most countries and patients themselves. In the United State, the newly diagnosed patients spent approximately $8941 more than subjects who were not diagnosed with DM over a period of 5 years . ApproxiA comprehensive search has been made on the PubMed, Embase, web of science, and the Cochrane Library databases to find all the eligible studies up to July 13th 2019. The following text words were used in the PubMed: (“diabetes” OR “glucose intolerance” OR “insulin resistance” OR “hyperglycemia” OR “hyperinsulinemia” OR “metabolic syndrome”) AND . Correlative key words were used in the Embase, web of secience, and the Cochrane Library. To comprehensively search eligible studies, we simultaneously searched the reference lists of relevant reviews or included publications for further studies.The included literatures met the following criteria: (1) cohort design; (2) investigated gallbladder cancer outcomes; (3) assessed the gallbladder cancer mortality with or without DM; (4) reported the information of hazard ratios (HRs), relative risks (RRs), or standard mortality ratios (SMRs). The exclusion criteria were as follows: (1) case-control or cross-sectional design; (2) unavailable data.Two authors independently extracted all data from publications using the same criteria. The following data were included: the first author’s name, publication year, country, sample size, the number of male or female participants, mean age at baseline, average follow-up duration, diabetes assessment, and adjusted factors.2 and Q statistics. Both I2 > 50% and P value< 0.1 were regarded as high heterogeneity. We conducted subgroup analysis to evaluate the potential sources of heterogeneity from country, follow-up duration, diabetes assessment, and adjusted factors . A sensitivity analysis was performed by removing each study from the overall analysis to investigate the influence of a single study. We used funnel plots, Begg and Egger tests to assess publication bias. P value< 0.05 was viewed as a significant level. The statistical analyses were performed with Stata software (version 12.0).We used Reviewer Manager 5.3 in this meta-analysis to analyze the data. The pooled HRs with 95% CIs were calculated as the effect estimates for the relationship between DM and gallbladder cancer mortality. The fixed-model was used when the heterogeneity was low, while the random-model was used when the heterogeneity was high. Owing to the low gallbladder cancer mortality in general population, the RRs, SMR were considered the similar estimates of the HRs . StatistDetailed study selection process was described in Fig. The baseline characteristics of the included studies were listed in Table The quality assessment results were shown in Figs. P < 0.00001; Fig. 2 = 0%; P = 0.95). The sensitivity analysis results indicated that the summary HR ranged from 1.09 (95%CI: 1.06–1.13) when excluding study from Chen 2017 to 1.12 (95%CI: 1.07–1.17) when excluding study from Currie 2012 [Eight studies focused on the relationship between diabetes mellitus and gallbladder cancer mortality. We merged the data of these studies and found that pre-existing diabetes had a high correlation with the mortality of gallbladder cancer compared with non-DM participants .A total of four studies estimated the difference of gallbladder cancer mortality between male and female DM patients. The analysis was conducted to see if female DM patients had a higher risk of gallbladder cancer mortality then male patients. The pooled analysis results demonstrated that no significant differences had existed between DM men and women  and Begg test (P = 0.845) showed no significant evidence of publication bias.The symmetric funnel plots indicated a potential low publication bias Fig. . MoreoveThis meta-analysis of cohort studies provided comprehensive evidence that the diabetes mellitus had an impact on the survival of patients with gallbladder cancer. Our results suggested that diabetes patients had a higher mortality rate of gallbladder cancer compared with non-diabetes patients. And the results were independent of country, follow-up duration, diabetes assessment, BMI, smoking, or education. Though previous analysis had indicated that DM women were more likely to develop gallbladder cancer than DM men due to sex hormones , we founSeveral physiological mechanisms might account for the increase of gallbladder cancer mortality in DM patients. A growing number of studies have found that overweight, obesity, metabolic syndrome, and insulin resistance were closely related to the increase of gallbladder disease –39. HypeTo our knowledge, our meta-analysis was the first study focused on the impact of DM on the survival of patients with gallbladder cancer. Previous study has proved that diabetes might increase the risk of gallbladder diseases . One metThe present meta-analyses had some strengths, including prospective design of cohort studies, eligible data from large sample size, detailed subgroup analyses, and low heterogeneity. Our findings provided an important message for patients with comorbid DM and gallbladder cancer that preventing the progression of diabetes might increase the survival from gallbladder cancer.There were several potential limitations in our study. First, residual confounding could not be ignored. Compared with non-DM participants, DM patients often had less healthy lifestyles, including higher rate of obesity, less physically activity, and more likely to smoke and drink. Though most of the included studies have adjusted these factors and our subgroup analysis showed no obvious heterogeneity between subgroups, we could not completely exclude the influence of these factors. Second, most studies did not tell the differences between type 1 and type 2 DM, though the majority of individuals were type 2 survivals. Older individuals were more likely to develop type 2 DM, while type 1 DM was a more common type in younger individuals. As a result of incomplete initial data on distinguishing this difference, some degree of inaccuracy of results was inevitable. Third, the number of eligible literatures remained low, which might have some influence on the final conclusion. The results of the difference of gallbladder cancer mortality between male and female patients remained open to question due to the lack of data and a high heterogeneity. Forth, the effect of medicine had not taken into account in the researches. Many studies have indicated that metformin, a commonly used diabetic medication, could retard the development of some cancers. None of the included researches have made adjustments for the use of diabetic medication. Last but not least, the multiplicity might exist in this analysis. The multiplicity attributed to a number of factors. On one hand, the subjects came from various backgrounds. Different rate, country, and age aggravated the multiplicity. On the other hand, the subjects from different studies might have an overlap.In conclusion, this meta-analysis suggested that diabetes patients had a higher mortality of gallbladder cancer. More relevant studies were needed to certify this association and tell the difference between men and women.


The summary RR per 5 kg/m2 increment in BMI at age 18 years was 1.17 , and per 10 cm increase in waist circumference was 1.13 . Higher BMI in middle age, BMI at age 18 years, and waist circumference were associated with increased rheumatoid arthritis risk, suggesting adiposity could be targeted for primary prevention.Several studies have investigated associations between overweight/obesity and risk of developing rheumatoid arthritis, however, the evidence is not entirely consistent, and previous meta-analyses mainly included case–control studies, which can be affected by various biases. We therefore conducted a systematic review and meta-analysis of cohort studies on adiposity and risk of rheumatoid arthritis. Relevant studies were identified by searching PubMed and Embase databases. Random effects models were used to estimate summary relative risks (RRs) and 95% confidence intervals (CIs) for rheumatoid arthritis in relation to different measures of adiposity. Thirteen cohort studies (10 publications) were included. The summary RR per 5 kg/m Globally an estimated 23.7 million people live with rheumatoid arthritis3. There is wide geographic variation in the prevalence rates of rheumatoid arthritis between countries, with rates ranging from 0.2 to 1.1% between regions, and higher rates in high-income countries compared to low- and middle-income countries5. For example, prevalence rates of 0–3 cases per 1,000 persons have been reported in some areas in Africa, while rates in Northern and Western Europe are around 4–11 cases per 1,000 inhabitants, and in North America between 9 and 11 cases per 1,000 inhabitants4. These geographical differences in the occurrence of rheumatoid arthritis may be attributed to various factors, including genetic factors, socioeconomic factors, access to health services, and lifestyle factors6.Rheumatoid arthritis is a systemic autoimmune disease characterized by an inflamed synovium which can cause pain, swelling, stiffness and deformity of multiple joints7, genetic predisposition (human leukocyte antigen (HLA)-DR4)8, smoking10, diet12, physical activity13, and obesity24.In addition, the global prevalence of rheumatoid arthritis has been reported to be approximately two times higher in women than in men25. These sex differences suggest that hormone-related factors or other sex-specific exposures may contribute to the development of rheumatoid arthritis26. Since body fat distribution differs for men and women, it is also possible that the differences in adiposity between men and women might, at least in part, account for the sex difference in rates of rheumatoid arthritis27.Established or suspected risk factors for rheumatoid arthritis include age19, but other studies found no clear association21. Although several meta-analyses have found positive associations between body mass index (BMI) and risk of rheumatoid arthritis24, they were mainly based on case–control studies which could be affected by recall and selection biases and reverse causation7. In addition, these meta-analyses only examined BMI and did not consider other measures of adiposity which might be more pertinent for rheumatoid arthritis. A few studies have suggested positive associations for BMI in early adulthood with rheumatoid arthritis15, and other studies suggested that abdominal fatness may also be important19, however, these results have not previously been summarised in a meta-analysis. Clarifying whether abdominal adiposity is more relevant for rheumatoid arthritis than general adiposity could contribute towards a better understanding of the underlying mechanisms and more targeted interventions for prevention and treatment. Therefore, to provide a more comprehensive assessment, we conducted an updated systematic review and meta-analysis of cohort studies to clarify the strength and shape of the association between different measures of adiposity and the risk of developing rheumatoid arthritis.Several epidemiological studies have suggested an increased risk of rheumatoid arthritis associated with overweight and obesityAn electronic literature search was conducted in Embase and PubMed databases to identify all eligible studies on the association between adiposity and the risk of rheumatoid arthritis that were published up to May 10th, 2019. The following search strategy was used in PubMed and a similar search was adapted for the Embase search:  AND (“rheumatoid arthritis” or rheumatoid arthritis[MeSH]) AND .Records identified from the search strategy were screened based on the title and abstract in the first step, and then further assessed for eligibility based on the full text in the second step. Prospective and retrospective cohort studies, case-cohort studies, and nested case–control studies within cohort studies, which reported adjusted relative risk (RR) estimates , incidence rate ratios, odds ratios (ORs)) and 95% confidence intervals (CIs) for rheumatoid arthritis for at least three or more categories of any measure of adiposity  or a risk estimate on a continuous scale were eligible for inclusion in this meta-analysis. Grey literature such as conference abstracts were excluded because they are unlikely to contain all the information required for dose–response analyses and for evaluation of study quality. Reviews, letters, comments, editorials, meta-analyses, meta-synthesis, prediction models, protocols, as well as epidemiological studies in patient populations were excluded. If there were duplicate publications from the same study population, the publication with the largest number of cases or with the most detailed information needed for dose–response meta-analyses was included. The initial screening of all references based on abstracts and titles was done by one author (TO) and a second author (DA) independently did the second part of the screening in duplicate.The following data were extracted from each publication: name of the first author, publication year, country, study name or description, study period, follow-up duration, number of participants or controls, number of cases, measurement of anthropometry , exposure , comparison (the contrast or metric of the exposure), RR, OR, HR and 95% CI, and adjustment for covariates. If several multivariable models were reported in the paper, the effect estimate most fully adjusted for potential confounders was used for the meta-analysis.28. Each study was appraised in terms of methodological quality categories: selection, comparability and outcome, with a total allowable score of up to 9 stars.Study quality of the included studies was assessed using the Newcastle–Ottawa scale2 increase in BMI and per 10 cm increase in waist circumference29. The average of the natural logarithm of the RRs was estimated and the RR from each study was weighted using random effects weights29. Given the low prevalence of rheumatoid arthritis and the moderate effect sizes observed, different effect measures  were not converted, but used directly. If studies reported results by sex or other subgroups, but not overall, the subgroup-specific RR estimates were pooled with a fixed effects model before inclusion in the overall meta-analysis. To explore the dose–response relationship between increasing levels of adiposity and the risk of developing rheumatoid arthritis, linear and nonlinear dose–response analyses were conducted. For the linear dose–response analyses, the method of Greenland and Longnecker was used31. Study-specific linear dose–response slopes were calculated in a logit-linear model using the estimates across at least three categories of adiposity or by using continuous estimates directly. Fractional polynomial models were used to investigate potential nonlinear associations32. The extent of heterogeneity was assessed using the I2 statistic which describes the percentage of variability across studies in a meta-analysis, and I2 values of approximately 25%, 50% and 75% indicated low, moderate and high heterogeneity, respectively33. Publication bias was assessed with funnel plots34, the Egger linear regression test35, and the Begg rank correlation test36. If there was evidence of publication bias as indicated by p < 0.10 for Egger’s or Begg’s test or by asymmetry in the funnel plot, a sensitivity analysis was conducted to evaluate the potential impact of publication bias on the summary estimates by excluding outlying studies in the funnel plot from the analysis. To ensure that the results were not driven by one large study or a study with an extreme result, sensitivity analyses were conducted sequentially excluding each study from the analysis and assessing its impact on the summary estimate. Subgroup analyses were conducted to evaluate whether the results differed according to the following characteristics: sex, number of cases , geographic location , measurement of BMI , self-reported ), duration of follow-up  and the extent of adjustment for confounding factors including age, education, social class, smoking, alcohol intake, physical activity, diabetes, and other comorbidities. All statistical analyses were performed using Stata/IC 13.1 .Random effects models were used to calculate summary RRs and 95% CIs per 5 kg/mFigure 38 and 5 studies (3 publications) were nested case–control studies within cohort studies39. Three publications included data from two studies each39, and one of these provided estimates from two different studies combined only19. While 8 studies (6 publications) included both men and women39, one study included men only37 and 4 studies (3 publications) were restricted to women only20. Eight studies (6 publications) performed anthropometric measures through a clinical examination39, while information on measures of adiposity was self-reported by participants in 4 studies (3 publications)20. In a nested case–control study using the UK General Practice Research Database (GPRD) conducted by Rodríguez et al.21, BMI was obtained from “physician-completed patient records”. To clarify this, we contacted the corresponding author of the study, as well as the Clinical Practice Research Datalink, who confirmed that although there is no way to be certain, it is reasonable to assume the anthropometric measurements were likely to have been performed by a physician or nurse at GP practices. Three studies  reported results on waist circumference19, and three studies (two publications) reported on BMI in early adulthood (at age 18 years)20. Only one study reported on waist-to-hip ratio20, and body fat percentage18, and no studies reported results for hip circumference, thus analyses of these exposures were not possible. Most studies adjusted for potential confounders such as age, sex, smoking, and alcohol consumption. The average study quality score using the Newcastle–Ottawa scale28 was 8, with seven studies scoring 9 stars39, one study scoring 8 stars21, four studies scoring 7 stars38, and one study scoring 6 stars20  showed some suggestion of publication bias, but not Egger's test (P = 0.14). When excluding one outlying study , the summary RR per 5 kg/m2 increase in BMI was 1.14 (95% CI 1.08–1.19), the heterogeneity was reduced  when the NHSII study by Lu et al.15 was excluded, to 1.14  when the study by Heliövaara et al.37 was excluded  20 were included in the analysis of BMI in early adulthood (at age 18 years) and risk of developing rheumatoid arthritis. The summary RR per 5 kg/m2 increase in BMI in early adulthood (at age 18 years) was 1.17  or Begg’s test (P = 0.57), although the number of studies was limited. There was no evidence of a nonlinear association between BMI in early adulthood (at age 18 years) and the risk of rheumatoid arthritis (Pnonlinearity = 0.99) 19 were included in the meta-analysis of waist circumference and risk of rheumatoid arthritis. The summary RR per 10 cm increase in waist circumference was 1.13 P = 0.02), with a positive association observed in women 1.15  based on data from 5 cohort studies and 6 case–control studies. The heterogeneity between studies was much lower in the current meta-analysis than in the previous meta-analysis, and this is probably to a large extent because we only included cohort studies. This avoids recall bias and reduces the possibility of selection bias and reverse causation. The larger number of cohort studies in the present analysis contributed to more precise risk estimates, and enabled assessment of associations of other adiposity measures, such as waist circumference and BMI in early adulthood, with risk of rheumatoid arthritis. Some studies have suggested that bariatric surgery may improve symptoms in patients with rheumatoid arthritis41, however, a recent Swedish study found no significant association between bariatric surgery and the risk of developing rheumatoid arthritis , but statistical power to detect a moderate association may have been low and confidence intervals were wide42.The current findings are consistent with a previous meta-analysis, which also found increased risk of rheumatoid arthritis with increasing BMI44. In a review by Colebatch et al.43, the authors summarised evidence on the effect of early life environmental factors on the risk of future rheumatoid arthritis. Among the findings, it was reported that early initiation of breastfeeding may be associated with a reduced risk of rheumatoid arthritis45, while high birth weight46 and maternal smoking in pregnancy47 may contribute to an increased risk of rheumatoid arthritis. On the other hand, high BMI in early adulthood is correlated with high BMI in middle age48, and it is possible that longer duration of obesity may confer an even greater risk of rheumatoid arthritis.The finding that high BMI in early adulthood (at age 18 years) was associated with an increased risk of rheumatoid arthritis might suggest that early life risk factors may be of importance in the aetiology of rheumatoid arthritis49. This is supported by evidence that adipocytes produce not only adipocytokines but also inflammatory cytokines49. Thus, adiposity can increase systemic inflammation, leading to increased susceptibility for developing rheumatoid arthritis.With regard to the underlying mechanisms that could contribute to the observed associations, it is known that adipose tissue plays an important role in the inflammatory process24, and we identified two additional cohort studies38 that were not identified in the most recently published meta-analysis in 201950. The inclusion of additional studies and a large total number of cases  and nearly 0.5 million participants provided sufficient statistical power to detect moderate associations and contributed to more robust results. Another major advantage of the current meta-analysis was that it only included prospective cohort studies. While the previous meta-analyses included many retrospective case–control studies that may have been affected by recall and selection biases50, the current meta-analysis is less likely to have been affected by such biases. Detailed dose–response analyses were conducted to clarify the dose–response relationship between adiposity and rheumatoid arthritis, and detailed subgroup analyses were conducted to investigate sources of heterogeneity. A novelty of this study was the analyses of additional measures of adiposity, specifically waist circumference and BMI in early adulthood, with risk of rheumatoid arthritis. BMI is considered to be a useful indicator of obesity, but this index does not reflect abdominal body fat distribution, and the accuracy of the index can vary according to age, sex and body composition51. When developing prevention interventions for rheumatoid arthritis, it may be informative to assess the association between several other measures of adiposity and the risk of developing rheumatoid arthritis.There are several strengths of this study. The current meta-analysis had 2–3 times as many cohort studies as three meta-analyses published in 2015, 2016 and 201839, and showed similar results to the overall summary estimate. Only four studies used self-reported data20. There was no heterogeneity in associations in subgroup analyses stratified by whether anthropometric measures were measured or self-reported.There are several limitations of this study that should be noted. In some of the included studies anthropometric measurements were self-reported. Consequently, the results might have been influenced by misclassification of the exposure. However, nine of the thirteen included studies performed anthropometric measurements through a clinical examination52 and other studies have found sugar-sweetened beverages to increase weight gain53. Given that little is known with certainty about diet and rheumatoid arthritis, and most studies have not adjusted for dietary factors, we cannot exclude the possibility that the observed association partly could be due to confounding by sugar-sweetened beverages or other dietary factors. Although there was some indication of publication bias in the analysis of BMI, this appeared to be explained by one outlying study, which when excluded did not substantially alter the results, but which also appeared to explain a large part of the heterogeneity between studies. Thus, it seems less likely that publication bias would have substantially impacted the overall conclusions. Another limitation was that the study populations were limited to North American and European populations and therefore had limited geographical coverage. Therefore, further research is needed to evaluate the association between adiposity and the risk of rheumatoid arthritis in other regions, including Africa, South America and Asia–Pacific.Heterogeneity between studies, publication bias and confounding are potentially important limitations of any meta-analysis. Some degree of heterogeneity is expected because studies may differ with regard to the assessment of adiposity, other risk factors, the outcome, underlying risk factor profile and disease rates, as well as analytical approaches including the extent of adjustment for confounding factors. Sources of heterogeneity were investigated in subgroup analyses, however, with the exception of the difference in association for men and women, there was little evidence of heterogeneity between other subgroups including geographic location, number of cases, measurement of BMI, duration of follow-up, and adjustment for confounding factors . Although most of the individual included studies adjusted for important potential confounders, it is important to note that residual and unmeasured confounding is possible. For example, one study suggested that consumption of sugar-sweetened beverages may increase risk of rheumatoid arthritis25. In addition, overweight and obesity have become a considerable problem in most regions of the world54, including low-income and middle-income countries such as China, Brazil and Indonesia as well as in high-income countries55.According to the Global Burden of Disease 2010 study, the number of people living with rheumatoid arthritis is expected to increase considerably over the coming decades, especially in low-income and middle-income countries56, it is possible that certain lifestyle-related risk factors  may contribute to the development of rheumatoid arthritis partly through an effect on adiposity59, and further studies are needed to clarify if this is the case.As the number of studies on adiposity variables other than BMI was limited, further studies are needed on a range of alternative adiposity measures including waist circumference, hip circumference, waist-to-hip ratio, waist-to-height ratio, BMI in early adulthood, and weight change throughout the life course, and to clarify whether the sex-specific results for BMI also apply to these other measures of adiposity. Because excess weight to a large degree is driven by diet and lifestyle factorsThis study confirms a positive association between levels of adiposity and the risk of developing rheumatoid arthritis. Higher BMI in middle age, BMI in early adulthood (at age 18 years), and waist circumference were associated with a higher risk of rheumatoid arthritis in dose–response meta-analyses. The current analysis suggests that excess weight may be a risk factor for rheumatoid arthritis in women, but further cohort studies are needed to clarify sex-specific associations with adiposity measures other than BMI as the available data is limited. Since most of the existing studies are from Europe and the US, additional studies are needed from other geographic regions as well. Based on these findings, prevention strategies targeting weight reduction and maintenance of a healthy weight may contribute to reducing the burden rheumatoid arthritis places on society.Supplementary information.


Dietary and lifestyle factors may play an important role in the increasing prevalence of nephrolithiasis. We aimed to review and quantify the associations between lifestyle factors and incident nephrolithiasis and suggest lifestyle changes for the primary prevention of nephrolithiasis.2 statistic was employed to evaluate heterogeneity. Subgroup analysis, sensitivity analysis and meta-regression were also conducted whenever possible.PubMed, EMBASE, and Cochrane Library were searched up to May 2019, for observational studies and randomized controlled trials (RCTs) that assessed modifiable lifestyle factors and risk of nephrolithiasis in adults. Pooled relative risks (RRs) and 95% confidence intervals (CIs) were computed using a random effects model. The IFifty relevant articles with 1,322,133 participants and 21,030 cases in total were identified. Prominent risk factors for incident stones were body mass index , dietary sodium , fructose, meat, animal protein, and soda. In contrast, protective factors included fluid intake , a Dietary Approaches to Stop Hypertension (DASH) style diet , alcohol , water, coffee, tea, vegetables, fruits, dietary fiber, dietary calcium , and potassium. Vitamin D  and calcium  supplementation alone increased the risk of stones in meta-analyses of observational studies, but not in RCTs, where the cosupplementation conferred significant risk.Several modifiable factors, notably fluid intake, dietary patterns, and obesity, were significantly associated with nephrolithiasis. Long-term RCTs are required to investigate the cost-effectiveness of dietary patterns for stone prevention. The independent and combined effects of vitamin D and calcium supplementation on nephrolithiasis need further elucidation. Nephrolithiasis, though usually manifested as a benign painful condition, poses a tremendous threat to the health care system and individual well-being. It incurs an enormous cost estimated at US $2.1 billion in 2000 and with a projected annual increase of US $1.24 billion by 2030 as a result of population growth and the pandemics of diabetes and obesity –10. WorsNevertheless, the prevalence and incidence of kidney stones have continued to grow globally, and despite the higher prevalence in males, up to 10.6% in 2007–2010, the gender gap appears to be closing due to a disproportionally increased prevalence in females –14. AlthWe, therefore, conducted a comprehensive systematic review and meta-analysis with an aim to build an evidence-based list of modifiable lifestyle and dietary factors and to establish corresponding lifestyle recommendations.Two authors (BBL and MEL) independently searched the databases, screened the articles retrieved, abstracted the data and assessed the quality of included studies. Discrepancies were resolved by consensus. This study followed MOOSE  and PRISMA  guidelines.PubMed, EMBASE and Cochrane Library were searched from inception to May 2019 using the terms “risk” AND “kidney stone” with limitation to publications in English. Full search terms and included studies are listed in Additional file Publications were considered eligible if they fulfilled the following inclusion criteria: i) observational studies or RCTs of adults without prior history of nephrolithiasis; ii) assessment of at least one modifiable risk factor; and iii) provided risk estimates with 95% CI, or enough information to calculate them. Exclusion criteria were also applied: i) population with recurrent or prevalent kidney stones as this condition might have affected their dietary behaviors; ii) studies featuring nonmodifiable risk factors, genetic or urinary lithogenic factors; and iii) cross-sectional studies or RCTs with short duration of follow-up  and small sample size (< 100 participants). Bibliographies of included studies and related reviews were hand-searched for additional publications. If multiple articles reported on the same cohorts, we only selected the most recent one unless different risk factors were assessed. When a single paper reported both individual and pooled risk estimates of different cohorts, or presented gender-specific results, we chose to analyze them individually in the meta-analysis.We recorded the first author’s last name, publication year, country, study design, study period, participants’ genders and baseline ages, the number of participants and cases, assessment factors and methods, exposure comparison, outcomes, maximally adjusted risk estimates with 95% CIs, and confounding factors by using piloted data extraction sheets. The incidence rate ratios (IRRs) and 95% CIs were inverted for one article  that use2 statistic, with values greater than 50 and 75% representing moderate and substantial heterogeneity, respectively. To identify sources of heterogeneity, subgroup and meta-regression analyses were performed according to study design, gender, region, study quality, number of cases, durations of follow-up and confounding factors whenever applicable. We used funnel plot, Egger’s test and the trim-and-fill method to evaluate publication bias and small study effects only if the number of included studies was ≥10. Sensitivity analyses were performed to evaluate each study’s effect on the overall risk estimates , beverage (n = 12), diet (n = 19), physical activity (n = 4), energy intake (n = 3), vitamins and calcium supplements (n = 22). All the included RCTs reported data available for secondary analysis of the associations with vitamin D and calcium supplementation.The selection process with exclusion reasons is presented in Fig. Dietary factors were assessed in 10 observational studies (19 publications), of which 8 were cohort studies conducted within the USA and Europe and 2 were case-controlled studies in the USA and China.In the meta-analysis of high vs low intake Fig. , a high As shown in Fig. 2 = 77.4%). The pooled result of 3 cohorts [Dietary calcium , potassium  and magnesium  intake were inversely associated with the risk of incident nephrolithiasis, whereas high intake of dietary sodium increased the risk by 38%  including 11 prospective cohort studies and 2 case-controlled studies investigated the effect of beverages on kidney stones, of which 5 were conducted within the USA, 4 in Europe, and 4 in Asia.2 = 79.3%) was reduced to I2 = 0.0%, while the association remained significant  after excluding 2 studies without adjustment for dietary factors [2 = 35.8%) after excluding one study [Total fluid intake up to 2 l/d reduced the risk of incident stones by almost half  when compared with less than 1 l/d  assessed the association with vitamin consumption and calcium supplement. The meta-analyses of observational studies and RCTs are summarized in Fig. No significant associations with incident stones were observed for high intake of dietary vitamin B6 and total vitamin D. Meta-analyses of total and supplemental vitamin C showed nonsignificantly increased risks of incident stones with moderate and substantial heterogeneity, respectively. The pooled results of 2 female cohorts , 36 for Seventeen studies in 15 articles examined the effect of obesity on incident kidney stones, with 16, 5 and 2 studies investigating BMI, waist circumference (WC) and waist-to-hip ratio (WHR), respectively. Of 17 studies, 7 were conducted in the Americas, 8 in Asia, and 2 in Europe.2 = 71.2%) and publication bias (P = 0.002) were present  existed.Physical activity was assessed in 4 articles with 7 cohort studies. Four studies were conducted in the USA, 2 in China, and 1 in Finland. The meta-analysis of the 7 studies revealed slightly but not significantly decreased risk of incident stones  with increased physical activity level  investigated the association between energy intake and nephrolithiasis. A nonsignificant 12% increase in the risk of incident stones was found with energy intake ≥2500 kcal/d   mg/d lower urine citrate, 24  mg/d higher urine uric acid, 0.13  lower urine pH, and higher relative supersaturation for uric acid . This, iAn inverse association of dietary calcium with kidney stones was found because high dietary calcium prevented oxalate absorption by forming calcium oxalate complex in the guts, thus reducing urinary oxalate excretion . HoweverA DASH-style diet, which contains a large number of fruits, vegetables and dairy products and low saturated fat , signifiIn our meta-analyses, increased fluid, alcohol and beer uptake decreased the risk of nephrolithiasis, followed by high intake of coffee, tea and water. In contrast, a high intake of soda promoted stone formation. One RCT demonstrated that urinary volume could be manipulated to reduce the recurrence rate by high water intake . It is mCalcium and vitamin D supplementation was controversial due to inconsistent meta-analyses of RCTs and observational studies. The NHS I  reportedPooled analysis of high vs low BMI yielded a significant 39% increased risk of incident stones. Insulin resistance, a feature of obesity  and typePrimary preventive measures, if well developed and implemented among the stone prone population, may limit the substantial negative impact of nephrolithiasis on the health care system and individuals. It was shown that primary preventive measures such as increased water intake could significantly decrease cost and stone burden in virtual cohorts . As the There are various limitations to our study. First, although a few articles reported that calcium-oxalate stones were the predominant type in small subsamples of NHS I (77%), NHS II (79%) and HPFS (86%) , most stThe summarized evidence in this study indicated that several lifestyle factors, such as BMI, beverages and dietary patterns, were significantly associated with the risk of nephrolithiasis. Adherence to specific dietary patterns is likely to have a far-reaching effect in reducing the stone burden. However, long-term RCTs are needed to confirm the cost-effectiveness of dietary patterns for the reduction of the incidence and recurrence rates of kidney stones. Future studies on vitamin D and calcium supplementation could help clarify their respective and combined effects on kidney stone formation.Additional file 1. Search strategies of PubMed, EMBASE, and Cochrane Library and the resulting list of included studies.Additional file 2. Summary of Newcastle Ottawa Scale scores for observational studies.Additional file 3. Risk of bias summary and graph for included randomized controlled trials in meta-analysis.Additional file 4. Sensitivity analyses of the included studies.Additional file 5. Characteristics of the included studies of incident kidney stones.Additional file 6. Funnel plot (A) and Egger’s test (B) for publication bias and small study effects among studies assessing the association between body mass index and risk of incident nephrolithiasis.Additional file 7. Subgroup meta-analyses and meta-regression of body mass index and incident kidney stones.


The risk factors contributing to embolism in cardiac myxoma (CM) are yet controversial. This systematic review and meta-analysis aimed to clarify the risk factors of embolism for the CM patients.PubMed, Embase, Cochrane library, Web of Science, China National Knowledge Infrastructure, Wan Fang, and Wei Pu databases were searched from inception to June 2019. Statistical analysis was conducted using Stata version 14.0. The pooled odds ratio or mean difference with 95% confidence interval was estimated for each risk factor.P < 0.01), hypertension (P = 0.03), irregular tumor surface (P < 0.01), tumor in atypical location (P = 0.01), narrow base of tumor (P < 0.01), and increased fibrinogen (FIB) (P < 0.01) are significant risk factors of embolism in CM patients. However, sex, age, body mass index, smoking, left ventricular ejection fraction, diabetes, hyperlipidemia, atrial fibrillation, valvular heart disease, coronary heart disease, tumor size, platelet count, white blood cells, and hemoglobin were not associated with embolism .Herein, 12 studies, encompassing 1814 patients, were included. The pooled results suggested that New York Heart Association (NYHA) class I/II (NYHA class (I/II), hypertension, irregular tumor surface, atypical tumor location, the narrow base of tumor, and increased FIB were significant risk factors of embolism in CM patients. For CM patients with these factors, early surgery might be beneficial to prevent embolism. Cardiac myxomas (CMs) are the most prevalent type of primary cardiac tumors in adults  that accAlthough, several risk factors of embolism are recently reported –16, someThe Preferred Reporting Items for Systematic Reviews and Meta-Analyses statement was followed . The PubThe studies that fulfilled the following criteria were eligible for inclusion in this meta-analysis: 1) All CM patients who underwent surgical resection and were confirmed by pathological examination; 2) Patients in each study were classified into two groups (embolic and non-embolic); 3) Embolic events were diagnosed by clinical symptoms or imaging examination. If the same population was presented in more than one publication, the study with the largest sample was included.We excluded the studies that met at least one of the following criteria: 1) Sample size < 50; 2) Abstracts from conferences, letters to the editor, and reviews; 3) Incomplete data.Two authors (L.P and J.W) reviewed all retrieved articles and extracted data independently. The titles and abstracts were first screened to identify potentially eligible articles, and then, full texts were read to confirm their eligibility for inclusion in this meta-analysis. The extracted data included the following information: first author, year of publication, country, study type, sample size, age, incidence of embolism, and risk factors that include patient characteristics  class and atrial fibrillation (AF)), tumor characteristics (tumor surface and tumor location), and hematological parameters (white blood cell (WBC) and platelet count (PLT)). The NYHA class was divided into two groups (I/II vs. III/IV). The tumor surface was classified as irregular and regular . The tumThe quality of the included studies was assessed using the Joanna Briggs Institute (JBI) Critical Appraisal Checklist  for Case2) test. I2 = 25% was considered low, 50% was moderate, and 75% was high. When I2 > 50% or P-value < 0.05 was identified for substantial heterogeneity, we used the random effect model; otherwise, a fixed-effect model was adopted. A sensitivity analysis was conducted when substantial heterogeneity was presented. Publication bias was assessed by the Egger’s test. All tests were two-sided, and the results were considered statistically significant at P < 0.05.Stata 14.0 was used for the statistical analysis of all data. The odds ratio/mean difference (OR/MD) and 95% confidence interval (CI) were calculated using the fixed-effect model or the random effect model. The heterogeneity among studies was quantified using the Cochran’s Q test and chi-square   , age , body mass index (BMI) , smoking , left ventricular ejection fraction (LVEF) , diabetes , hyperlipidemia , AF , valvular heart disease , and coronary heart disease .A meta-analysis was performed to assess the 12 risk factors, of which NYHA class (I/II)     was not associated with the condition.A meta-analysis was performed for 4 risk factors. Of these, irregular tumor surface  Fig.  significP < 0.01) , WBC , and hemoglobin (HB) .A meta-analysis was performed for 4 risk factors. Of these, increased fibrinogen (FIB)  Fig.  signific2 = 76.8%), BMI (I2 = 83.7%), and coronary heart disease (I2 = 84.2%). Sensitivity analysis showed a similar trend among various studies except for the study by Yin et al. [2 = 0.00%, all P > 0.05); also, the overall effect did not differ significantly .High heterogeneity across studies was detected regarding age , and sensitivity analysis showed a similar trend among studies except for the study by Kalçık et al. [2 = 58.6%, P = 0.03), while the overall effect did not differ significantly (P = 0.19).In addition, tumor size had high heterogeneity  and tumor surface (P = 0.051).General considerations suggest that use of the Egger’s test with substantially fewer 10 studies would be unwise ; hence, In the current meta-analysis, a total of 20 potential factors were studied. The results suggested that the NYHA class, hypertension, tumor surface, tumor location, tumor base size, and FIB were independent risk factors associated with embolism.Among patient characteristics, we found that 2/12 factors (NYHA class and hypertension) were significantly associated with embolism. The NYHA class is the most commonly used indicator of clinical response to cardiac function. The current study showed that better cardiac function (class I/II) was at a significantly higher risk of embolism than the worse function (class III/IV). This finding might be attributed to the fact that most embolic group patients were diagnosed after an acute embolic event; also, the tumor blockage symptoms were relatively less, and the overall cardiac function was better than that of non-embolic group. Conversely, the non-embolic group patients have a prolonged course of the disease and are likely to show hemodynamic changes; this phenomenon was associated with obstructive heart failure. However, the LVEF was not associated with embolism. We think this difference may attribute to the fact that some patients exist heart failure with preserved ejection fraction, and they also could reach NYHA class III/IV . Hence, Regarding tumor characteristics, we found that three factors are related to embolism: tumor surface, location, and base. Macroscopically, the surface of the tumor is classified into two types , 28 as fIn terms of hematological parameters, the present study designated increased FIB as a risk factor of embolism. It is a substrate for thrombin and is directly involved in the clotting process. In addition, FIB promotes the aggregation of PLT, increases blood viscosity, and aggravates red blood cell adhesion, all of which promote thrombosis . Thus, wTo the best of our knowledge, this is the first meta-analysis to investigate the risk factors of embolism in patients with CM, with some practical clinical implications. Nevertheless, this meta-analysis also has some limitations. First, since all enrolled studies were retrospective case series, residual confounders and unidentified factors were inevitable in observational studies. Second, all the included studies originated from Asia, which might lead to selection bias. Third, the number of overlapped risk factors in each of the studies was small for a comprehensive statistical analysis. Finally, with respect to the tumor size, the data of our meta-analysis showed high heterogeneity. Although one major source of heterogeneity was detected by sensitivity analysis, other differences between the studies should be considered.This systematic review and meta-analysis identified the following significant risk factors of embolism for CM patients: NYHA class I/II, hypertension, irregular tumor surface, atypical tumor location, the narrow base of tumor, and increased FIB. For CM patients with these factors, early surgery may be beneficial for preventing embolism.Additional file 1.Additional file 2.Additional file 3.


In conclusion, the study shows for the first time a robust and inverse association between white meat consumption and all-cause mortality and a neutral association with CV mortality and morbidity. This highlights the importance of differentiating the meat types for what concerns their health effects and suggests that white meat might be a healthier alternative to read and processed meat consumption.The association of meat consumption with mortality and morbidity for non-communicable diseases has been extensively studied. However, the relation of white meat consumption with health outcomes remains controversial. The present meta-analysis was conducted to comprehensively analyze the available evidence on the consistency and strength of the association between the consumption of white meat, death from any cause and incidence of fatal and non-fatal cardiovascular (CV) events. PubMed, Web of Science, Scopus and Embase databases were searched for articles published up to April 30, 2020. We included prospective cohort studies reporting relative risks and pertinent 95% confidence intervals (CI) for all-cause mortality and/or CV events . A total of 22 studies were included in the meta-analysis. Eleven studies (14 data sets) reported data on all-cause mortality, 10 studies (15 datasets) on cardiovascular disease (CVD) mortality and 10 studies (11 datasets) on non-fatal CV events. When comparing the highest versus the lowest consumption of white meat, the pooled OR and pertinent 95% CI were 0.94 (0.90, 0.97,  The relationship between dietary habits and chronic non-communicable diseases (NCDs) has been extensively investigated ,3,4,5,6.A systematic research of all the prospective cohort studies published until April 30, 2020 on the association of white meat consumption with all-cause mortality, incidence of cardiovascular disease and coronary heart disease, was performed according to PRISMA  guidelines .The research was carried out in the electronic databases  using the following key words in all possible combinations: (“poultry” or “white meat” or “chicken” or “turkey” or “rabbit”) and  and (“meat” or “consumption” or “intake” or “serving”). The search was limited to human studies and had no language restrictions.We included studies defining “white meat” as poultry  and rabbit. Exclusion criteria were: (i) retrospective studies; (ii) studies conducted on vegetarian people; and (iii) studies in which it was not possible to evaluate white meat alone because part of combined meals. When several publications of the same study were identified, only the most recent, or most detailed publication was used.https://www.crd.york.ac.uk/PROSPERO (accessed on 26 January 2021)) with the registration number: CRD42020198126.The reference lists of selected studies and reviews were also searched to identify additional articles not previously included. In addition, the lists of the retrieved articles were manually reviewed. In case of missing data, the Authors were contacted by e-mail to acquire the original data. This meta-analysis has been registered on Prospero (Two reviewers (I.C. and A.G.) assessed the titles and abstracts of all identified studies and independently reviewed and extracted relevant data from each study, including first author and year of publication, country, study design, sample size, participant characteristics, follow-up duration, number of subjects in comparison groups, type of white meat, and main results of the outcomes investigated. In case of disagreement, a third investigator was consulted (M.V.). Discrepancies were resolved by consensus.The evaluation of methodological quality of each study was performed with the Newcastle-Ottawa scale (NOS), which is specifically developed to assess quality of non-randomized observational studies . The scop < 0.05 being considered statically significant.The primary outcome was all-cause mortality defined as death for any cause (vascular and non-vascular). Secondary outcomes were CV mortality and non-fatal CV events . Data synthesis and statistical analysis were carried out using comprehensive meta-analysis  software. The pooled probability of all-cause mortality, cardiovascular mortality and cardiovascular events in subjects in the highest versus the lowest category of white meat consumption was expressed as OR with pertinent 95% confidence intervals (CI). The overall effect was tested by Z-scores, with 2 statistic, which measure the inconsistency across study results and describe the proportion of total variation in study estimates due to heterogeneity rather than sampling error. In detail, an I2 value of 25% corresponds to low, 25–50% to moderate, and 50% to high heterogeneity. Publication bias was assessed by the Egger’s test and funnel plots of the log OR versus the standard error were used as graphical representation.Statistical heterogeneity among studies was assessed by chi square Cochran’s Q test and with Ip < 0.10 being considered statistically significant. In case of a significant publication bias, the Duval and Tweedie’s trim and fill method with the random-effect model was used to allow for the estimation of an adjusted effect size. In order to be as conservative as possible, the random-effect method was used to take into account both the within-study and between-study variability.To address possible small-study effect, funnel plots were visually inspected for asymmetry and the Egger’s test was used to assess publication bias, over and above any subjective evaluation, with Separate analyses were performed by stratifying results for geographical area where each study was carried out. Analyses were repeated after excluding studies judged as low quality according to NOS score.We performed meta-regression analyses in order to assess whether differences in the risk of all-cause mortality, CV mortality and CV events observed between highest and lowest white meat consumption categories were confounded by demographic  or clinical variables  or follow-up duration. The regression models were carried out using differences in the risk of all-cause mortality, CV mortality and CV events observed between highest and lowest white meat consumption categories as the dependent variables (y) and the above-mentioned co-variates as independent variables (x). Comprehensive meta-analysis software  was used for the multivariate approach.After excluding duplicates, the search retrieved 1355 articles. Of these, 1291 were excluded because they were off the topic after scanning the title and/or the abstract; 42 because they were reviews/comments/case reports or lacked data of interest or did not match the inclusion and exclusion criteria. More in detail, among the studies excluded based on study design are also the study of Sinha et al. , becauseAmong the included studies Lee et al. , Nagao ep < 0.001; Of the 22 studies included in the meta-analysis, 11 studies reported data on all-cause mortality ,43,44,462: 95.6%, p < 0.001) and it was not reduced by the exclusion of one study at a time. Results were consistently confirmed when specifically analyzing the 4 studies (5 datasets) from America [p = 0.04; I2: 98.5%, p < 0.001; p = 0.01; I2: 12.9%, p = 0.33; p = 0.08; I2: 57.3%, p = 0.05; Heterogeneity among studies was significant  ,42,43,462: 82.1%, p < 0.001) and was not reduced by the exclusion of one study at a time. As shown in p = 0.12; I2: 93.8%, p < 0.001), Asia [p = 0.90; I2: 24.7%, p = 0.24) or Europe [p = 0.17; I2: 0%, p = 0.94).Heterogeneity among studies was significant (I1), Asia ,35,41,46r Europe ,42,43  reported data on non-fatal CV events ,38,44,452: 69.8%, p < 0.001) and was not reduced by the exclusion of one study at a time. Results were confirmed when separately analyzing studies from America  or Europe , while a significant risk reduction emerged from the Asian studies [p = 0.01; I2: 0%, p = 0.55).Heterogeneity among studies was significant , CV mortality  and non-fatal CV events .After excluding low quality studies , the results were entirely confirmed for all-cause mortality  was observed for studies on CVD events. These results were confirmed by Duval and Tweedie’s trim and fill analysis.Funnel plot examination  suggesteRegression models showed that none of the clinical and demographic characteristics evaluated impacted on the difference in all-cause mortality between subjects in the highest versus lowest white meat consumption categories . No metaDiet is one of the major modifiable factors that affect disease risk, thus it is of the greatest importance to identify dietary habits that decrease the risk of disease and death. The relationship of red and processed meat consumption with increased risk of all-cause death and incidence of CVD has been consistently demonstrated ,41,42,46The results have shown a 6% significantly lower all-cause mortality for subjects in the highest vs. the lowest white meat consumption category and an overall neutral association with CV mortality and morbidity. The findings for all cause death and CV death were fairly consistent when the analyses were stratified by geographical area  to partly account for different background diet and unmeasured lifestyle-related factors, including food preparation techniques. As for non-fatal CV events, the Asian studies indicated a significant risk reduction at variance with those from America and Europe. These should be further investigated due to the low number of meta-analyzed studies. There were not enough studies to perform the analyses by gender; however, the meta-regression analyses indicated that the clinical and demographic characteristics of the participants, including gender, did not impact on the difference in all-cause mortality.In a previous meta-analysis of six studies, one of which also included fish in the white meat definition, Abete et al. have shown a weak and inverse association between white meat consumption and all-cause death in women only, and no relationship with CV mortality . These cThe interpretation of the effects of white meat consumption on health is a difficult task, as subjects consuming more white meat are, at the same time, consuming less red meat. On the other hand, people with a low red meat intake may prefer others protein sources, such as proteins from vegetable origin that could have per se a beneficial impact on cardiovascular health. However, it is important to underline that white meat, unlike products of vegetable origin, is a source of high-quality proteins and may therefore fully substitute red meat.Furthermore, the background diet and the food preparation techniques, associated with the regular consumption of the different types of meat may also have a role which is difficult to account for. There are, however, plausible mechanisms which may partly mediate a different association of processed and red meat or white meat with health outcomes . Meats aLast, but not least, the study findings need to be evaluated also in the light of the growing awareness on the importance of the foods and beverages we produce, choose and consume in relation to their environmental sustainability ,51. It hThis study has some strengths. First, it includes a large number of studies and participants , thus allowing sufficient statistical power, second it provides data on non-fatal CV events which were not available before, furthermore, the analysis by geographical area confers internal consistency to the results by partly adjusting for unmeasured cultural and lifestyle related factors including background diet and procedures of food preparation.The potential limitations need also to be discussed. First, we must acknowledge that studies included in this meta-analysis have different inclusion and exclusion criteria and therefore subjects with different clinical and demographic characteristics were considered in the overall analysis. Moreover, meta-analyses are performed on aggregate data and the multivariate approach allows for the adjustment for some, but not all, potential confounders. Thus, although meta-regression models were used to refine our analyses by assessing the influence of most clinical and demographic variables on the observed results, caution is necessary in the interpretation of results due to potential residual confounding. Finally, our results were affected by a significant heterogeneity. Although it was not possible to definitively establish sources of such heterogeneity, all findings were substantially confirmed by appropriate sensitivity and subgroup analyses and the impact of clinical and demographic variables on the study results was evaluated by means of meta-regression models. Furthermore, we excluded the presence of publication bias by using different methods, and in case of significant bias our results were confirmed after trimming and imputing studies. In addition, we found heterogeneity among studies in the definition of white meat intake in the lowest and the highest consumption group  and in some studies there was missing information regarding type of white meat . Moreover, the potential confounding of cooking methods and preparation techniques could not be assessed.In conclusion, the study shows a robust and inverse association between the consumption of unprocessed white meat and all-cause mortality, and a neutral association with CV mortality and morbidity. These findings highlight the importance of differentiating the meat types and suggest that white meat might be a ‘healthy’ and environmentally more sustainable alternative to red and processed meat consumption. Most people consume meats on a regular basis and therefore, notwithstanding the relatively modest risk reduction (6%), the findings of this study have relevant public health implications and provide evidence to inform dietary guidelines for a healthy and environmentally sustainable nutrition.


Q-test combined with the inconsistency index (I2) was used to check the heterogeneity between studies. A total of six case-control studies on rs2910164 and rs11614913, 4 studies on rs3746444 and rs11134527, and three studies on rs531564 were included. No evidence of association was found between miR-146a rs2910164, miR-196a2 rs11614913, miRNA-499 rs3746444, and miR-218 rs11134527 polymorphisms and cervical cancer risk in all the genetic models. The miR-124 rs531564 polymorphism was associated with a statistically increased risk of cervical cancer in a homozygote model , dominant model , and recessive model . However, this finding should be interpreted with caution for limited samples and heterogeneity. Large-scale and well-designed studies are needed to validate our result.The objective of this study was to conduct a meta-analysis to systematically summarize and investigate the association of miRNA-124 rs531564, miRNA-218 rs11134527, miRNA-146a rs2910164, miRNA-196a2 rs11614913, and miRNA-499 rs3746444 polymorphisms with cervical cancer. A systematic review was performed to identify relevant studies using Embase and PubMed databases. A chi-square-based  Cervical cancer (CC) is the fourth most common cancer in females, with 570,000 new cases and 311,000 deaths estimated for 2018 worldwide [MicroRNAs (miRNAs) are short nonprotein-coding small RNAs of approximately 23 nucleotides that mediate posttranscriptional regulation through base pairing to partially complementary sites of the target mRNA and promote translational repression or messenger RNA degradation , 7. IncrRecently, though a number of studies have been conducted to investigate the association of miR-124 G>C (rs531564), miR-218 A>G rs11134527), miR-146a G>C (rs2910164), miR-196a2 T>C (rs11614913), and miR-499 A>G (rs3746444) polymorphisms with cervical cancer 134527, m, the resP value. The selection of the study was completed independently by two investigators (Liu and Dong) according to the inclusion criteria.The PubMed and Embase databases were searched by two independent investigators covering all papers published up to December 27, 2020. The following terms were used: “cervical or CIN”, “microRNA or miRNA or microRNAs”, and “polymorphism or polymorphisms”. The references of the retrieved articles and review articles were also searched manually for further relevant studies. All eligible studies met the following criteria: (1) full-text study, (2) case-control design, (3) analysis of the relationship between miRNA polymorphisms and the risk of cervical cancer, (4) studies focusing on human beings, and (5) sufficient data to calculate risk estimates: the odds ratio (OR) with 95% confidence interval (CI) and a Each publication was sought for the following information: the first author's surname, the publication year, country of origin, ethnicity, genotyping method, the number of cases and controls, and Hardy-Weinberg equilibrium (HWE). According to the amount of the sample, we defined sample size as the large sample size (≥200) and small sample size (<200).Two reviewers (Liu and Dong) extracted the eligible research data repeatedly. Disagreement between two reviewers was discussed with another reviewer (Wang) until a consensus is reached.χ2 test was used to evaluate the HWE of polymorphism in the control group. If P < 0.05, it is considered deviated from HWE.When Hardy-Weinberg equilibrium (HWE) in the control was not reported, the The quality of the selected studies was assessed according to the methodological quality assessment scale which was mentioned in prior meta-analysis  see Tab.In this scale, five items were evaluated, including (1) the representativeness of the subjects, (2) the ascertainment of cervical cancer, (3) the control of the genotyping methods, (4) HWE in controls, and (5) assessment of association. The quality score ranged from 0 to 9, and a high score indicated good quality of the study. Studies with a score less than 5 were removed from the subsequent analyses.We conducted our meta-analysis based on the PRISMA checklists and followed the guideline . The strQ-test combined with the inconsistency index (I2) was used to check the heterogeneity between studies [P < 0.1 or I2 ≥ 50%) [P value of Egger's test <0.05 was considered representative of statistically significant publication bias. All statistical analyses were carried out using the STATA 12.0 software .A chi-square-based  studies . The ran2 ≥ 50%) , 27. BesA total of 247 studies were acquired from Embase and PubMed databases (Embase: 98 and PubMed: 149). The detailed screening process is shown in In six studies, genotype frequencies of rs531564, rs11134527, rs2910164, rs11614913, and rs3746444 were presented separately, so each of them was considered as a separate study in this meta-analysis. Therefore, all the six included studies containing 2291 cases and 2850 controls for rs2910164, six studies containing 2291 cases and 2850 controls for rs11614913, four studies involving 1554 cases and 1962 controls for rs3746444, four studies containing 3167 cases and 3132 controls for rs11134527, and three studies containing 975 cases and 950 controls for rs531564 were finally analyzed in our meta-analysis. For these five miRNA in this meta-analysis, the subjects in all included studies were of the Asian population.The TaqMan assay and polymerase chain reaction-restriction fragment length polymorphism (PCR-RFLP) were used for genotyping to determine the SNPs in the included studies. HWE of genotype distribution in the controls was tested in all studies, and two articles were not consistent with HWE. And according to the methodology quality assessment, all the studies were above a score of 6.0 and recruited into the following analyses.PH = 0.001, PH = 0.140; CC vs. GG: OR =0.85, 95% CI 0.57-1.26, PH = 0.001; GC/GG vs. CC: OR = 0.90, 95% CI 0.70-1.15, PH = 0.014; and GG vs. GC/CC: OR = 0.90, 95% CI 0.68-1.19, PH = 0.002).The association strength between miRNA-146a rs2910164 polymorphism and cervical cancer risk is shown in PH = 0.800), while there is a significantly decreased risk of cervical cancer susceptibility in heterozygote comparison , homozygote model , dominant model , and recessive model .Next, subgroup analysis was conducted according to sample size and HWE. In the large sample size and HWE group, a significant association was also not identified in any genetic model. And the heterogeneity is still high. When we delete three articles , heterogeneity is significantly reduced . A signiPH = 0.000, PH = 0.001; CC vs. TT: OR = 141, 95% CI: 0.94-2.10, PH = 0.000; TC/CC vs. TT: OR = 1.28, 95% CI: 0.93-1.75, PH = 0.000; and CC vs. TC/TT: OR = 1.23, 95% CI: 0.97-1.29, PH = 0.021) when all eligible studies were pooled.Significant association was not identified in any genetic model . In the HWE group, heterogeneity is also significantly reduced. But significant association was not identified in any genetic model.Next, subgroup analysis was conducted according to sample size and HWE. In large sample size, heterogeneity is significantly reduced. A significantly increased risk of cervical cancer susceptibility was identified in homozygote comparison .Significant association was not identified in any genetic model , heterogeneity is significantly reduced in all models . HoweverPH = 0.003; AG vs. AA: OR = 1.07, 95% CI: 0.89-1.29, PH = 0.053, PH = 0.006; AG/GG vs. AA: OR = 1.06, 95% CI: 0.86-1.31, PH = 0.091; and GG vs. AG/AA: OR = 0.97, 95% CI: 0.76-1.23, PH = 0.035).Significant association was not identified in any genetic model , while there is a significantly decreased risk of cervical cancer susceptibility in the homozygote model  and recessive model . Significant association was not identified in heterozygote comparison  and dominant model .The sample size of the 4 included articles is all large size and subgroup analysis was conducted according to HWE. In the HWE group, heterogeneity is also significantly reduced. But significant association was not identified in any genetic model. When we delete two articles (Li Chuanyin and Guange Chen), heterogeneity is significantly reduced in all models . We obsePH = 0.887, PH = 0.409), and recessive model , while there is a significantly decreased risk of cervical cancer susceptibility in allelic comparison . Significant association was not identified in heterozygote comparison .We observed a significantly increased risk of cervical cancer susceptibility in the homozygote model  and recessive model , while there is a significantly decreased risk of cervical cancer susceptibility in allelic comparison . Significant association was not identified in heterozygote comparison  and dominant model .Next, subgroup analysis was conducted according to sample size. In a large sample size, a significantly increased risk of cervical cancer susceptibility could be identified in the homozygote model , miR-196a2 (rs11614913), miR-499 (rs3746444), and miR-218(rs11134527), we conducted a subgroup analysis by sample size, and the results showed that the sample size is the source of heterogeneity of miR-196a2 (rs11614913) in each genetic model. However, for other SNPs, heterogeneity was still high in large sample size. However, when we delete some articles, we found that the heterogeneity is significantly reduced.The influence of each study on the pooled ORs in this meta-analysis was examined by excluding one study at a time using STATA 12.0 software. The results showed that no significant alteration in the pooled ORs was found in any of the genetic models for the five SNPs.Due to the limited number of studies included (<10), no funnel plot or Egger's test was performed.In this meta-analysis, the relationship between these five SNPs in microRNAs  and the risk of cervical cancer was evaluated from 11 published studies. We demonstrated that no evidence was found for the association between miRNA-146a rs2910164, miR-196a2 rs11614913, miRNA-499 rs3746444, and miR-218 rs11134527 and cervical cancer risk in any genetic models. The miR-124 rs531564 polymorphism was associated with a statistically increased risk of cervical cancer in the homozygote model, dominant model, and recessive model. However, due to limited samples and heterogeneity, this finding should be interpreted with caution.κB signaling pathway mediated by IRAK1 and TRAF6, which is an important signaling factor involved in the occurrence of cancers [Recent studies have demonstrated that miRNAs are abnormally expressed in cervical cancer and play a role in tumor generation . miR-146 cancers , 33. Pre cancers , 35. Evi cancers .β3 (LAMB3) has been verified as a transcriptional target of miR-218 [In 2013, Yamamoto et al. found that the expression of miR-218 in cervical cancer tissues was significantly lower than that of adjacent noncancerous tissues. The restoration of miR-218 significantly inhibited the proliferation, migration, and invasion of cancer cells . Laminin miR-218 , and the miR-218 . The miR miR-218 . Results miR-218 . miR-499 miR-218 , 42.Significant heterogeneity was found in this meta-analysis for association of miR-218 rs11134527, miR-146a rs2910164, miR-196a2 rs11614913, miR-499 rs3746444, and cervical cancer. Therefore, we conducted a subgroup analysis based on sample size and HWE. When we restricted the sample size to the large sample size group and the HWE group, heterogeneity is significantly reduced for miR-196a2 rs11614913. However, the small sample size group of miR-196a2 includes only two studies, and the large sample size group includes three studies. Therefore, further research is needed to confirm these results. Significant heterogeneity was also found for the association of miRNA-146a rs2910164, miR-499 rs3746444, miR-218 rs11134527, and cervical cancer in the large sample size group and the HWE group. However, when we delete some researches in each analysis, heterogeneity is significantly reduced, suggesting that these researches contributed to the source of heterogeneity. However, owing to the limited articles included, further evaluation of the results is still needed.In the sensitivity analysis, no significant changes were found after omitting one study at a time, indicating that our meta-analysis results are relatively stable and credible.Our meta-analysis has several advantages. First, this is the first meta-analysis of the association between five miRNA polymorphisms and the susceptibility to cervical cancer. In addition, according to the methodological quality assessment, all included studies are of high quality. Moreover, there are no restrictions in the literature search, so the selection bias is well controlled. Our meta-analysis has some limitations. Firstly, the number of included studies is too few. Secondly, heterogeneity was detected in miR-146a rs2910164, miR-196a2 rs11614913, miR-499 rs3746444, and miR-218 rs11134527. After deleting some studies in each analysis, heterogeneity is significantly reduced. The pooled ORs became significant without evidence of heterogeneity. Thirdly, the included studies were all Asian ethnicity; therefore, it is uncertain whether these results can be generalized to other populations. Fourth, publication bias might also exist, because only published studies were included in this meta-analysis, and results that are not statistically significant usually have less chance of publication.In summary, our results suggested that miR-124 rs531564 polymorphism is significantly associated with increased risk of cervical cancer, while miR-146a rs2910164, miR-196a2 rs11614913, miR-499 rs3746444, and miR-218 rs2292832 polymorphisms may not be associated with the susceptibility of cervical cancer. However, since few studies have been included, there is not enough data to fully confirm the association between cervical cancer and the five miRNA polymorphisms, so the results should be interpreted with caution.


Background: ABCG2 and ABCB1 are genes related to the pharmacokinetics of sunitinib and have been associated with its toxicity and efficacy. However, the results have been controversial. This study aimed to evaluate the associations of ABCG2 and ABCB1 polymorphisms with sunitinib-induced toxicity and efficacy in renal cell carcinoma (RCC) by meta-analysis.Methods:PubMed, EMBASE, Cochrane Library, and Web of Science were systematically searched for studies investigating the associations of the ABCG2 rs2231142 polymorphism with sunitinib-induced toxicity and the associations of the ABCB1 rs1128503 and ABCB1 rs2032582 polymorphisms with sunitinib-induced toxicity and clinical outcomes. The associations were evaluated by effect size (ES) with 95% confidence intervals (CIs).Results: Eight and five studies were included in the toxicity and efficacy analysis, respectively, including a total of 1081 RCC patients. The ABCG2 rs2231142 A allele was associated with an increased risk of sunitinib-induced thrombocytopenia and hand-foot syndrome (HFS) in Asians . However, the ABCG2 rs2231142 polymorphism was not associated with sunitinib-induced hypertension or neutropenia . Compared with the C allele, the ABCB1 rs1128503 T allele was associated with a decreased risk of sunitinib-induced hypertension but worse progression-free survival (PFS) . There was no significant association between the T allele or C allele of ABCB1 rs1128503 and overall survival (OS) . The ABCB1 rs2032582 T allele was associated with worse PFS than the other alleles , while there was no significant association between the T allele or other alleles and sunitinib-induced hypertension, HFS, or OS .Conclusion: The results indicate that the ABCG2 rs2231142 polymorphism may serve as a predictor of sunitinib-induced thrombocytopenia and HFS in Asians, while the ABCB1 rs1128503 polymorphism may serve as a predictor of sunitinib-induced hypertension, and both the ABCB1 rs1128503 and rs2032582 polymorphisms may serve as predictors of PFS in RCC. These results suggest a possible application of individualized use of sunitinib according to the genetic background of patients. Renal cell carcinoma (RCC) is a common cancer with high malignancy, and 20–30% of RCC patients already suffer from metastatic lesions at the time of diagnosis . SunitinSunitinib is a substrate of ATP-binding cassette member B1 (ABCB1) and another efflux transporter encoded by ATP-binding cassette member G2 (ABCG2) . ABCG2 aIdentifying the effects of ABCG2 and ABCB1 polymorphisms on sunitinib-induced toxicity and efficacy in patients with RCC could help to optimize the therapeutic management strategy and maximize the clinical benefits of sunitinib. Here we used a meta-analysis to evaluate the associations of ABCG2 rs2231142, ABCB1 rs1128503, and ABCB1 rs2032582 polymorphisms with sunitinib-induced toxicity and those of ABCB1 rs1128503 and ABCB1 rs2032582 polymorphisms with PFS and OS in patients with RCC.PubMed, EMBASE, Cochrane Library, and Web of Science were systematically searched using the keywords “sunitinib’’ and ‘‘polymorphisms” to identify eligible studies. Only articles in English were included. The last search was updated on 7 February 2020. We also retrieved eligible studies in the references of all relevant articles. Furthermore, we contact the authors to request missing information to strengthen the analysis.The inclusion criteria for studies were as follows: 1) studies of RCC patients who received sunitinib treatment; 2) studies evaluating the associations of ABCG2 rs2231142 polymorphism with sunitinib-induced thrombocytopenia, HFS, and hypertension; 3) studies evaluating the associations of ABCB1 rs1128503 and rs2032582 polymorphisms with sunitinib-induced hypertension and HFS; 4) studies evaluating the associations of ABCB1 rs1128503 and rs2032582 polymorphisms with PFS and OS; and 5) studies with available odds ratios (ORs) between genotypes.The exclusion criteria for studies were as follows: 1) letters, reviews, and case reports; 2) nonclinical studies; and 3) studies that were duplicate publications. In addition, we pooled complete data in the meta-analysis if multiple studies had overlapping data.Two authors independently selected the relevant articles and then extracted the following data from the articles: first author’s name, publication year, study design, sample size, treatment regimen, age, sex, region/ethnicity, ABCG2 rs2231142, ABCB1 rs1128503 and ABCB1 rs2032582 phenotype, genotyping methods, evidence of Hardy-Weinberg equilibrium in cases, adverse events and clinical outcome in ABCG2 rs2231142, ABCB1 rs1128503 and ABCB1 rs2032582 phenotypes, and the ORs of ABCG2 rs2231142 A allele vs. C allele, ABCB1 rs1128503 T allele vs. C allele, and ABCB1 rs2032582 T allele vs. other alleles. If there was any controversy, it was resolved by discussion among the authors.2 tests. Both no-heterogeneity criteria were required to be met (p > 0.05 AND I2 < 50%) to use the fixed-effects model; otherwise, the random-effects model was used to calculate the pooled ES. The Z test was used to investigate the significance of the ES, and PZ < 0.05 was considered statistically significant. The stability of the results was evaluated by sensitivity analysis. The publication bias among studies was determined by Egger’s and Begg’s tests, and PE < 0.05 was considered significant.STATA software  was used to perform the statistical tests. The associations of ABCG2 rs2231142, ABCB1 rs1128503, and ABCB1 rs2032582 polymorphisms with sunitinib-induced toxicity, and ABCB1 rs1128503 and ABCB1 rs2032582 polymorphisms with PFS and OS were evaluated by effect size (ES) with 95% CIs . The hetPubMed, EMBASE, Science Direct, and Web of Science. We included 34 relevant studies after removing the duplicates and reviewing the titles, abstracts, and full texts. Then, 25 articles were excluded because they did not determine the ORs between genotypes. Ultimately, 8 articles published between 2011 and 2018 assessing the relationships of ABCG2 rs2231142, ABCB1 rs1128503, and ABCB1 rs2032582 polymorphisms with sunitinib-induced toxicity, and 5 articles published between 2011 and 2017 assessing the relationships of ABCB1 rs1128503 and ABCB1 rs2032582 polymorphisms with PFS and OS were included in the current meta-analysis , and the fixed-effects model was applied to the analysis. There was heterogeneity between studies assessing the relationships of ABCG2 rs2231142 polymorphism with sunitinib-induced hypertension, HFS, and neutropenia , and the random-effects model was applied to the analysis. Based on the results, we found that compared with the C allele, the ABCG2 rs2231142 A allele was significantly associated with an increased risk of sunitinib-induced thrombocytopenia in Asians , and the fixed-effects model was applied to the analysis. There was heterogeneity between studies assessing the relationship of ABCB1 rs1128503 polymorphism with sunitinib-induced PFS and OS , and the random-effects model was applied to the analysis. The results showed that compared with the C allele, the ABCB1 rs1128503 T allele was significantly associated with a decreased risk of sunitinib-induced hypertension and worse PFS , and the random-effects model was applied to the analysis. There was no heterogeneity between studies assessing the relationship of the ABCB1 rs2032582 polymorphism with sunitinib-induced HFS or PFS , and the fixed-effects model was applied to the analysis. The results showed that compared with the other alleles, the ABCB1 rs2032582 T allele was significantly associated with worse PFS  and OS  were not changed, and no single study qualitatively altered the pooled ES in either analysis  decrease in the risk of neutropenia and inferior PFS and OS  1, 2, and 3; Fms-like tyrosine kinase 3 receptor (FLT3); and PK-related genes, including cytochrome P450 1A1 (CYP1A1), CYP3A5, NR1/2, and NR1/3, have also been reported to be associated with sunitinib-induced toxicity and efficacy in patients with mRCC . SNPs rsWe did not include rs1045642 in ABCB1 in the meta-analysis because we could not obtain the ORs between the genotypes or because the included article numbers were below three. ABCG2 polymorphisms have also been reported to be associated with other sunitinib-induced toxicity in mRCC patients, such as hypothyroidism . The T aTo our knowledge, the present study is the first meta-analysis to investigate the associations of ABCB1 and ABCG2 polymorphisms with sunitinib-induced toxicity and efficacy in patients with RCC. Based on our results, compared with the C allele, the ABCG2 rs2231142 allele A was significantly associated with increased risks of sunitinib-induced thrombocytopenia and HFS in Asians, while the T alleles of ABCB1 rs1128503 and ABCB1 rs2032582 were significantly associated with worse PFS than the other alleles. The results indicate that the ABCG2 rs2231142 polymorphism may serve as a predictor of sunitinib-induced thrombocytopenia and HFS in Asians, while the ABCB1 rs1128503 polymorphism may serve as a predictor of sunitinib-induced hypertension, and both the ABCB1 rs1128503 and rs2032582 polymorphisms may serve as predictors of PFS in RCC. Although our analysis showed no significant relationships of ABCB1 and ABCG2 polymorphisms with other sunitinib-induced severe toxicity or OS, we should also pay more attention to the use of sunitinib in Asian patients. The results may support the possible application of individualized use of sunitinib according to the genetic background of patients with RCC. Genotyping for ABCG2 rs2231142, ABCB1 rs1128503 and rs2032582 polymorphism could become a clinical routine practice to select the appropriate dose to decrease the risk of sunitinib-induced thrombocytopenia and HFS in Asians while ensuring efficacy. The population pharmacokinetic model based on the genotypes should be established to predict the occurrence of sunitinib-induced toxicities and the efficacy.