Presently, whether X-ray repair cross complementing group 3 (XRCC3) Thr241Met polymorphism is correlated to leukemia risk remains controversial. Because of this reason, the objective of current study is to explore whether XRCC3 Thr241Met polymorphism confers risk to leukemia.Two independent authors systematically and comprehensively searched Pubmed, Embase, the Cochrane library, Google academic, China National Knowledge Infrastructure (CNKI). Search time is from database foundation to March 2021.Overall, significant associations between leukemia risk and XRCC3 Thr241Met polymorphism were found in Caucasian population by allele contrast , homozygote comparison , and recessive genetic model .The present meta-analysis suggests that the XRCC3 Thr241Met polymorphism may be a risk factor for leukemia in Caucasian population.The online version contains supplementary material available at 10.1186/s12920-021-01076-w. Leukemia is a very frequent malignance tumor originating from hematopoietic stem cells. The leukemia cell stops at different stages of cell development due to uncontrolled proliferation, dysdifferentiation and aleukemiatosis block. Its common symptoms are anemia, infection and bleeding. The incidence of leukemia in China is approximately 3–8 individuals per 100,000 . About OSeveral studies have shown that the occurrence of leukemia is associated with exposure to risk environment factors such as benzene, formaldehyde, smoking history, residence decoration and the use of different kinds of hair dye. Benzene and its metabolites make bone marrow damaged by immune-mediated responses, leading to the occurrence of leukemia . A largeThe occurrence and development of leukemia is a complicated process. Many scholars attribute it to some risk factors including physical factors, chemical factors, and virus infection; however, these factors are not acting as a necessary role for leukemia occurrence and progression. Approximate 30% patients do not embrace physical factors, chemical factors, and virus infection will acquire leukemia. All the above evidences indicate that extra genetic or non-genetic factors modulating leukemia susceptibility are yet to be identified.Although leukemia pathogenesis is an extremely complicated process and the exact pathogenesis of leukemia is still unknown, studies have shown that DNA damage is closely related to its occurrence and development . Many reYan et al. published a literature in 2014 that also investigates the association between XRCC3 Thr241Met polymorphism and leukemia risk . RegrettAs far as we know, this is the first meta-analysis which comprehensively explores the association between XRCC3 Thr241Met polymorphism and leukemia susceptibility. The objective of current study is to estimate whether XRCC3 Thr241Met polymorphism confers risk to leukemia.https://www.ncbi.nlm.nih.gov/pubmed/), EMBASE (https://www.embase.com/), the Cochrane library(https://www.cochranelibrary.com/), Google academic (https://scholar.google.com/), and Chinese national knowledge internet (https:// www.cnki.net/). Search time is from database foundation to March 2021. The keywords applied in the search process were as follows: (“XRCC3” or “X-ray repair cross complementing group 3”) together with (“leukemia”). The literature language was limited to English language and Chinese language. Additionally, in order to avoid the omission of relevant literatures, we searched the references as much as possible.Two independent authors systematically and comprehensively searched Pubmed (The inclusion criteria must meet a series of conditions: (a) a case–control study; (b) making an assessment of the association between XRCC3 Thr241Met polymorphism and leukemia risk; (c) offering sufficient information and data to count OR and 95%CI. The exclusion criteria also must meet a few conditions: (a) patients with other hematological system diseases such as multiple myeloma, aplastic anemia, myelodysplastic syndrome, autoimmune hemolytic anemia, idiopathic thrombocytopenic purpura. (b) patients with some inflammatory diseases or cardiovascular and cerebrovascular diseases such as urinary tract infection or shock, acute myocardial infarction or unstable angina, rheumatoid arthritis or systemic lupus erythematosus. (c) not offering sufficient data for meta-analysis. (d) its experiment objective was pig, rat or other animals.All the necessary information was independently reviewed and assessed by first author and second author . Then this contradictory data or information was reassessed by the third author (Qiuhua Li). The extracted data consisted of author name, publication year, genotyping methods, sample size, ethnicity, matching criteria, source of control, HWE conformity. If the similar opinion could not reach in the course of data extraction, suggestion was offered by another experienced researcher (Qiuhua Li) to determine the correct selection. The similar method was applied equally to evaluation of literature quality. In the present meta-analysis, we applied the risk assessment criteria of Newcastle–Ottawa Scale (NOS) bias to evaluate the quality of each literature. The main criteria consisted of three aspects including selection of enrolled study subjects (0–4 scores); between-group comparability (0–2 scores); exposure outcomes and factors (0–3 scores). It should be noted that the ethics approval of our study was waived by Ethics Committee of The Fifth Affiliated Hospital of Zunyi Medical University as no human or animal was directly enrolled in our study and meta-analysis is the statistical analysis of large collection of analysis results from individual studies for the purpose of integrating the findings.2 statistics would be applied [2 < 50% was considered to low heterogeneity, 50 ≤ I2 < 75% was considered to moderate heterogeneity and I2 ≥ 75%was considered to significant heterogeneity. If I2 < 50% and P > 0.1, the fixed-effects model would be used. If I2 ≥ 50% or P ≤ 0.1, the random -effects model would be used. Furthermore, the Galbraith plot was used to spot the outliers to find out the potential heterogeneity as much as possible. Sensitive analysis was applied to detect the influential studies which might contribute obvious bias to final results. The funnel plot and Egger’s test were put into use to recognize the existence of publication bias [The association power was assessed through the corresponding indexes including OR and 95%CI. And both the Q-statistic and I applied . Four ge applied , 15. I2 ion bias . Meta-reion bias , 18. FurPRISMA 2009 Flow Diagram shows the flow chart of meta-analysis search course  has been subject to widespread attention as comprehensive modifications have been detected in mRNAs of human and yeast . It contEliminating the source of bias is of vital importance for gene polymorphism association meta-analysis. Hence, we have attempted to conduct all the three patterns in the present meta-analysis. Firstly, allele contrast was used to find out the high risk or low risk allele. Secondly, homozygote comparison was used to find out the high risk or low risk genotype. The last pattern is comparing homozygote genotype versus allele carriers. In the present study, the moderate-significant heterogeneity between studies occurred in the overall population. Common reasons for heterogeneity consist of differences in the investigated populations or in genotyping methods or in sample size or it may be derived from other risk factors. By performing meta-regression, and subgroup analysis, we found that ethnicity might contribute substantial heterogeneity to final results. By Galbraith plot analysis, we found the studies Liu et al. and Hamdy et al. were not within reasonable limits. Then we explored the two studies carefully and discovered their shortcomings. The P value < 0.05 of HWE in control group was found in literature of Liu et al. And we found the sample size < 100 participants in literature of Hamdy et al. The results of our meta-analysis were not altered by omitting the two studies. The results of sensitive analysis and publication bias demonstrated that the results of our meta-analysis were stable and reliable.Although the present meta-analysis is comprehensive and rigorous, there are still some disadvantages existing. Firstly, more studies with different ethnicities are also needed because different ethnicities have different genetic backgrounds. Various ethnicities should be investigated and discussed including African population, Asian population, mixed population and Caucasian population. Secondly, different kinds of confounding factors such as age, gender and radiation exposure are not taken into consideration due to limited dataset , 45. TheThe present meta-analysis suggests that the XRCC3 is a candidate gene for leukemia susceptibility. The XRCC3 Thr241Met polymorphism may be risk factor for leukemia in Caucasian population. Further studies investigating other confirmed genetic factors and possible gene–gene and gene-environmental interactions for XRCC3 Thr241Met polymorphism should be performed.Additional file 1: Table S1. PRISMA 2009 checklistAdditional file 2: Table S2. PRISMA 2009 Flow Diagram.Additional file 3: Figure S3. Results of evaluating publication bias of other four genetic models .


Gliomas are the most common primary intracranial tumors in adults. Inappropriate dietary habits are thought to be a risk factor for most human cancer, and glioma is no exception. However, the effect of dietary factors on glioma is not clear.This review aims to quantitatively evaluate the association between various dietary intakes and glioma using a meta-analysis.We searched articles on PubMed, the Cochrane Library, the Web of Science, and EMBASE from their inception until October 11, 2021. According to heterogeneity, the fixed-effects or random-effects model was selected to obtain the relative risk (RR) of merger. Based on the methods described by Greenland and Longnecker, we explored the dose-response relationship between dietary intakes and the risk of glioma. Subgroup analysis, sensitivity analysis, and publication bias were also used.RR = 0.82, 95%CI:0.71–0.93), total vegetables , green vegetables , and orange vegetables  significantly reduced the risk of glioma, while the highest intakes of grains , processed meats , and processed fish  significantly increased the risk of glioma. The results of subgroup and sensitivity analyses remained unchanged. In the dose-response relationship, only tea was statistically significant. Taking an extra cup of tea every day reduced the risk of glioma by 4%.This study reviewed 33 articles, including 3,606,015 controls and 8,831 patients with glioma. This study included 12 food groups. Compared with the lowest intakes, the highest intakes of tea (Our analysis suggests that the intakes of tea, total vegetables, green vegetables, and orange vegetables may reduce the risk of glioma, while the intakes of grains, processed meats, and processed fish may increase the risk of glioma. Therefore, the effect of dietary factors on glioma should not be ignored.Systematic Review Registration:https://www.crd.york.ac.uk/prospero/, CRD42022296658. Gliomas are the most prevalent types of adult brain tumors, accounting for 78% of malignant brain tumors . BecausePtrend− = 0.005; α-carotene: OR = 0.5, Ptrend− = 0.01, β-carotene: OR = 0.5, Ptrend− = 0.01)  and insufficient intakes of vegetables and fruits are thought to be a risk factor for most human cancer, and glioma is no exception . Current = 0.01) . Blowers = 0.01) . None of = 0.01) –12, but  = 0.01) . A popul = 0.01) . Similar = 0.01) , 16. The = 0.01) , 18. Pra = 0.01) . However = 0.01) . FinallyTo provide a quantitative assessment of the effect of dietary intakes on glioma risk, we synthesized all published observational studies on dietary factors and glioma. We used the dose-response meta-analysis to quantify this association between dietary factors and glioma risk and to determine whether the relationship is linear, targeting to reach some evidence for dietary factors to prevent glioma.Two authors independently conducted an extensive search of the Cochrane Library, PubMed, Web of Science, and Embase until October 11, 2021. The Cochrane Library search terms used for the title, abstract, and keywords were “glioma” OR “brain cancer” OR “brain tumor” combined with “diet” OR “food” OR “lifestyle” OR “nutrition” OR “fruit” OR “vegetable” OR “meat” OR “coffee” OR “tea” OR “fish” OR “vitamin.” The same retrieval strategy was also applied to the other databases. No document type, language, or other relevant restrictions were used in the retrieval process, and the unpublished articles were excluded. Two reviewers screened the titles and abstracts to select the articles and reviewed the full text. Any disagreements between the two authors were settled by a third author. In addition, we searched the references of the published meta-analysis to identify other potential articles.For the meta-analysis, we included the articles that met the following criteria: (1) Exposure: the dietary intakes of participants. The exposure of interest was dietary intakes of participants. The studies gave the daily intakes of each food or the overall intakes of a food group through food frequency questionnaires or dietary recall; (2) Outcome: glioma; and (3) Population: 18 years old and above. This was due to the large dietary differences between minors and adults.The exclusion criteria of the meta-analysis were as follows: (1) study population included minors (<18 years of age); (2) non-observational study ; (3) lacking effect size and 95%CI; and (4) if multiple studies used the data from the same population, a study with the largest sample size was included in this meta-analysis.For the articles that conformed to the inclusion criteria, the data in the articles were extracted independently by two authors according to the predesigned format. The extracted data included the first author, year of publication, country, study population, study type, age, sex, sample size, number of cases, dietary intake level, effect size, and 95%CI extracted from the most adjusted model. In case of disagreement during data extraction, the conflict would be submitted to a third author for adjudication.Each study was evaluated by two authors and handed over to a third party for adjudication in case of disagreement. As the included articles were observational studies, the Newcastle–Ottawa scale (NOS) was used to evaluate the quality of the study and the possible risk of bias .I2 statistics. If the heterogeneity was not statistically significant (I2 <50% and p > 0.10), the fixed-effects model was used to combine the effect size and 95%CI. Otherwise, the random-effects model was used. We conducted a subgroup analysis to determine whether the heterogeneity of the study came from the study type (case-control study and cohort study), the study population , and the study quality , to explore the potential sources of heterogeneity. We performed a sensitivity analysis. We successively omitted one study at a time to assess each study's relative impact on the total effect size estimation. Different statistical models (fixed-effects and random-effects model) were used to estimate effect size. Egger's test and Begg's test were used to detect publication bias.For the current meta-analysis, we conducted a meta-analysis based on the effect size and 95%CI between the highest quantile and the lowest or reference quantile of dietary intakes. Heterogeneity between studies was assessed by Q-value was applied to assess between-study heterogeneity.Subsequently, for dietary factors with significant results in the analysis of extreme categories, we also explored the dose-response relationship between dietary intakes and glioma risk. The method developed by Greenland and Longnecker was used to analyze the dose-response relationship in this study . For thip < 0.05 was considered as statistically significant.Stata 14.0 was used for all statistical analyses. Unless otherwise noted, I2= 23.2%, Pforheterogeneity = 0.230)        , total v= 0.029) , green v= 0.823) , and ora= 0.422)  signific= 0.281) , process= 0.019) , and pro= 0.896)  signific= 0.896) .RR = 0.81, 95%CI: 0.71–0.93). Total vegetables were statistically significant in the case-control subgroup . The heterogeneity of total vegetables in the subgroup analysis of study type decreased from 53.4 to 11.0%. For the study population, tea was statistically significant in the American population , total vegetables were statistically significant in the European population  and other populations . Grains were statistically significant in the American population  and European population . For study quality, tea and processed meats were statistically significant in the subgroup with the study quality score >7 . Total vegetables and grains were statistically significant in the subgroup with the study quality score ≤ 7 . The heterogeneity of processed meats in the subgroup analysis of study quality decreased from 46.4 to 37.4% . Similarly, excluding one grain study . It was speculated that these two studies might be the main reason for the heterogeneity of related dietary factors and glioma risk results.The impact of the individual study on the summary RR was assessed by repeating the meta-analysis after removing each study in turn . For teain study , the ovep-value of publication bias of all dietary intakes was >0.1, suggesting that the difference was not statistically significant, so the publication bias was not obvious in this study . Although total vegetables, orange vegetables, and processed meats had similar linear trends, the results were not significant due to an insufficient number of studies.Due to the limited number of available articles, only tea, total vegetables, orange vegetables, and processed meats could be analyzed for the dose-response relationship from 10 articles. The dose-response relationship between various dietary factors and the risk of glioma was shown in Based on 33 observational studies on dietary factors and glioma published from 1986 to 2021, involving 3,606,015 controls and 8,831 patients, our meta-analysis results showed that higher intakes of tea, total vegetables, green vegetables, and orange vegetables could significantly reduce the risk of glioma, while higher intakes of processed meats, grains, and processed fish could significantly increase the risk of glioma. However, coffee, red meats, fresh fish, fruits, and citrus fruits had no significant effect on the risk of glioma. For the results of the dose-response relationship, increasing the intake of a cup of tea per day could reduce the risk of glioma by 4%. Although there was a similar linear trend between total vegetables, orange vegetables, and processed meats and the risk of glioma, and the results were not significant. This may be due to the availability of a limited number of articles that investigate the dose-response relationship of these dietary factors on glioma.We conducted a subgroup analysis of dietary factors with significant results due to the study population, study type, and study quality to explore the sources of heterogeneity. Although the overall heterogeneity of vegetables and gliomas was not high, it was found through a subgroup analysis that the difference of study types may be the main source of heterogeneity. Seven of the nine studies were case-control studies. Recall bias resulting from dietary survey methods in the case-control studies must be considered. There were only two cohort studies, which limit the significance of the combined results to a certain extent. The heterogeneity of processed meats mainly came from the difference in the study quality. The articles with a score of <7 still had 34.7% heterogeneity after merging. In addition, the exploration of influencing factors of glioma in the Burch study focused more on environmental exposures, such as smoking, water source, and x-ray exposure. It did not adopt appropriate investigation methods to systematically investigate the dietary intakes of the subjects but only provided a small number of dietary data such as tea and processed meats. Moreover, the cases in this study were not newly diagnosed, and the recall bias was large. These factors may explain the heterogeneity in the studies involving tea and processed meats. Similarly, Blowers et al. also contributed most of the heterogeneity to the meta-analysis of cereals and gliomas. The study had a small sample size of only 94 cases. The heterogeneity of green vegetables, orange vegetables, and processed fish was very small. Because the number of studies was small, there was no subgroup analysis. In the sensitivity analysis, the dietary factors with significant results were consistent in the two models. No publication bias was found in all studies, which suggested that the research results were relatively stable.RR = 0.80, 95%CI: 0.62–0.98)     = 0.69, 95%CI: 0.51–0.94] and had the same effect on glioblastoma , which was consistent with our results . For a l62–0.98) . These a62–0.98) , 53. In 25–0.92) . So far,25–0.92) . Animal 25–0.92) , 57. Exp25–0.92) . However9, 1.79) . However9, 1.79) . Some st . Moreove05–0.84) , 65. Alt05–0.84) , and oth05–0.84) , 18. Due05–0.84) , which wTo date, our study is the largest meta-analysis with the largest sample size that investigates dietary factors and glioma. This study has some advantages. First, this study is the first meta-analysis involving the impact of multiple dietary factors on the incidence of glioma, including 12 food groups. It is the first study to find the impact of grains, green vegetables, and orange vegetables on the incidence risk of glioma through a meta-analysis. Although the number of articles is relatively small, it provides a new scientific basis for the prevention of glioma. Secondly, the included articles were strictly screened. We rejected some studies that had been included in previous meta-analyses but involved meningiomas and other brain tumors. The source of heterogeneity is thoroughly analyzed for the significant results. All these improve the accuracy of the research results. However, this study has some limitations. First, the articles used have the publication dates that span a larger timeframe. Some long-standing articles and case-control studies have problems with quality, such as large recall bias and inaccurate dietary survey methods. These will inevitably impact the analysis. Additionally, as a meta-analysis of observational studies, we may not be able to control all potential confounding factors. These confounding factors may lead to a certain level of deviation. Finally, for the dose-response relationship, only tea obtained significant results. Other dietary factors did not obtain meaningful results due to a small number of articles that were included in the analysis. In future research, we can improve the relevant analysis by supplementing more articles.In summary, the current meta-analysis shows that the intakes of tea, total vegetables, green vegetables, and orange vegetables may reduce the risk of glioma, while the intakes of grains, processed meats, and processed fish may increase the risk of glioma. Therefore, the impact of dietary factors on glioma could not be ignored. In future studies, we should find a method to comprehensively evaluate dietary factors and nutrient exposure in the population, to further study the relationship between dietary factors and glioma.The original contributions presented in the study are included in the article/WL and WZ contributed to the conception or design of the work. WZ and JJ contributed to searching the databases. WZ, JJ, and YH contributed to the acquisition, analysis, or interpretation of data for the work. WZ and XL performed proofreading and modified the language. XL and FC reviewed and edited the manuscript. All authors have read and approved the final manuscript.This study was supported by the National Natural Science Foundation of Beijing (No. J200003) and National Science and Technology Major Project of China (No. 2016ZX09101017).The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


Background: The association between isolated diastolic hypertension (IDH) and cardiovascular events has been inconsistently reported. This meta-analysis of cohort studies was designed to investigate the effect of the 2018 European Society of Cardiology (ESC) definition of IDH on the risk of composite cardiovascular events, cardiovascular mortality, all-cause mortality, and all strokes including ischemic stroke (IS) and hemorrhagic stroke (HS).Methods: PubMed, Embase, the Cochrane Library, and Web of Science were searched from inception to July 6, 2021. Cohort studies that investigated the association between IDH and cardiovascular events risk, compared to normotension, were included. Pooled hazard ratios (HRs) and 95% CIs were calculated using a random-effects models and heterogeneity was evaluated using Q-test and I2 statistic. The robustness of the associations was identified using sensitivity analysis. The methodological quality of the studies was assessed using the Newcastle–Ottawa scale. Publication bias was assessed using funnel plot, trim-and-fill method, Begg's test, and Egger's test.Results: A total of 15 cohort studies (13 articles) including 489,814 participants were included in this meta-analysis. The follow-up period ranged from 4.3 to 29 years. IDH was significantly associated with an increased risk of composite cardiovascular events , cardiovascular mortality , all strokes , and HS , but not associated with all-cause mortality  and IS . Subgroup analysis further indicated that IDH in the younger patients (mean age ≤ 55 years) and from Asia were significantly associated with an increased risk of composite cardiovascular events, while the elderly patients (mean age ≥ 55 years), Americans, and Europeans were not significantly associated with an increased risk of composite cardiovascular events.Conclusion: This meta-analysis provides evidence that IDH defined using the 2018 ESC criterion is significantly associated with an increased risk of composite cardiovascular events, cardiovascular mortality, all strokes and HS, but not significantly associated with all-cause death and IS. These findings also emphasize the importance for patients with IDH to have their blood pressure within normal, especially in the young adults and Asians.Trial Registration: PROSPERO, Identifier: CRD42021254108. Isolated diastolic hypertension (IDH) is an important subtype of hypertension defined as a systolic blood pressure (SBP) of < 130 mm Hg and a diastolic blood pressure (DBP) of at least 80 mm Hg according to the 2017 American College of Cardiology (ACC)/American Heart Association (AHA) criterion , and 5.2A recent prospective cohort study by Wu et al. indicated that IDH was associated with cerebral hemorrhage, myocardial infarction (MI), and total CVD compared to normotension . ConversThe meta-analysis was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA)  and the Studies were included if they met the following inclusion criteria: (1) cohort study; (2) performed in the general adult population (age > 18 years); (3) reported the associations of IDH with the composite of cardiovascular events, cardiovascular mortality, all-cause mortality, all-strokes, IS, and HS; (4) defined IDH and normotension based on the 2018 ESC guidelines (SBP < 140 mm Hg/DBP ≥ 90 mm Hg vs. SBP < 140 mm Hg/DBP < 90 mm Hg); and (5) reported hazard ratios (HRs) with corresponding 95% CIs for the association between IDH and cardiovascular events or sufficient data for their calculation. We excluded studies if they: (1) involved pregnant, critically ill, or hospitalized participants or (2) were published as comments, conference abstracts, or letters to the editor. When republished studies that included participants from the same cohort and reported similar outcome measures were found, articles reporting the most relevant data were selected. However, if duplicate studies provided information on different outcomes, they were included in the specific outcome analysis. Two investigators (MYH and HQ) independently screened all the titles or abstracts and reviewed the full texts to determine the eligibility of the identified studies and the validity of the extracted data. Any disagreements were resolved through a discussion or by a third reviewer (CGF).Two investigators (MYH and LZL) independently extracted data from each eligible publication using a standardized data collection form. Any disagreements were resolved by consulting a third investigator (CGF). We used HRs to measure the associations. The primary outcomes of interest in this study were the composite of cardiovascular events. The secondary outcomes of interest were cardiovascular mortality, all-cause mortality, all strokes, IS, and HS. When an article was unavailable or to obtain additional information for analyses, an e-mail requesting the article or information was sent to the corresponding author. We recorded the following study characteristics: first author, publication year, study design, country, age at entry, percentage of male participants, cohort sample size, key exclusion criteria, outcomes, follow-up duration, treatment status at baseline, methods of BP measurement, ascertainment of outcomes, HRs and 95% CIs, and confounding variables adjusted in the multivariate analysis.Two investigators (MYH and HQ) assessed the study quality using the Newcastle–Ottawa quality assessment scale for cohort studies . This scp < 0.10) to assess the heterogeneity among studies and I2 statistic to quantify the percentage of the total variation due to that heterogeneity. Low, moderate, and high heterogeneity were defined as I2 values of 0–25, 26–75, and > 75%, respectively , irrelevant IDH definition (n = 3), improper comparison (n = 6), irrelevant normotension definition (n = 6), republished studies (n = 3), studies not reporting HRs (n = 2), abstract-only articles (n = 5), cross-sectional studies (n = 1), and reviews (n = 1). Finally, 13 articles with 489,814 participants were included in the meta-analysis, as one publications included three independent cohort studies , smoking (n = 14), sex (n = 13), BMI (n = 11), diabetes mellitus (n = 10), alcohol consumption (n = 9), antihypertensive treatment (n = 7), hypercholesterolemia (n = 7), education (n = 5), SBP (n = 5), race (n = 4), physical exercise (n = 3), and other relative confounders. The results of the quality assessment based on the Newcastle–Ottawa Scale are shown in ducation , previoup = 0.006) compared to normotension . In sensitivity analyses, the summary HRs ranged from 1.17 (95% CI: 1.04–1.32) to 1.34 (95% CI: 1.12–1.60) when individual studies were excluded from the analysis (p = 0.903) or Begg's test (p = 0.536) was found , with moderate heterogeneity  across the studies (p = 0.504) or Begg's test (p = 0.210) and no asymmetry was observed in the funnel plots , with moderate heterogeneity . Visual inspection of the funnel plot indicated mild asymmetry. This was further confirmed by a significant Egger's test (p = 0.048), while the p-values of Begg's test were statistically non-significant (p = 0.230). Two missing studies were imputed in the contour-enhanced funnel plots and the application of the trim-and-fill method did not change the risk estimate   .p = 0.03). Three studies with 108,327 participants assessed the association between IDH and IS (p = 0.137). Three studies with 108,327 participants evaluated the association between IDH and HS (p = 0.003).Four studies with 260,158 participants evaluated the association between IDH and all strokes , demonstH and IS , but notH and HS , indicatI2 = 0, Pheterogeneity = 0.362), participants from America  and Europe . Meta-regression analysis showed significant correlations between methods of BP measurement (p < 0.001). The source of heterogeneity among studies on the composite cardiovascular events may be due to the methods of BP measurement. The heterogeneity disappeared in groups when BP was measured in mercury sphygmomanometers  and automatic digital BP monitor (I2 < 0.1% Pheterogeneity = 0.583). Therefore, the source of heterogeneity among studies on the composite cardiovascular events may be due to the methods of BP measurement.Subgroup analyses stratified by mean age at entry, location, treatment status at baseline, BMI, and method of BP measurement were performed. IDH was significantly associated with an increased risk of composite cardiovascular events in most subgroups , except To the best of our knowledge, this is the first quantitative meta-analysis investigating the associations between the 2018 ESC definition of IDH and composite cardiovascular events, cardiovascular mortality, all-cause mortality, all strokes, IS, and HS. In this meta-analysis with 15 cohort studies involving 489,814 participants, three main findings emerged. First, the pooled results indicated that IDH is associated with an increased risk of composite cardiovascular events, cardiovascular mortality, all strokes, and HS, but not for all-cause mortality and IS. Second, IDH in the younger patients (mean age ≤ 55 years) was associated with an increased risk of composite cardiovascular events, but not in the elderly patients (mean age ≥ 55 years). Third, patients with IDH in Asia were significantly associated with an increased risk of composite cardiovascular events, while Americans and Europeans were not significantly associated with an increased risk of composite cardiovascular events.The previous meta-analysis showed that IDH defined according to the 2017 ACC/AHA criterion was not consistently associated with new-onset CVD and the relative size of any potential association was slight . While iFurthermore, we performed subgroup analyses stratified by age, BMI, geographic location, treatment status at baseline, and methods of BP measurement. Some significant findings were obtained. The relationship between IDH and the risk of cardiovascular events varies according to age, geographic location, and treatment status at baseline. In our subgroup analyses by location, a significant association between IDH and an increased risk of composite cardiovascular events was found in Asia, mainly in China. However, no statistically significant association was found between American and European populations. This was an interesting finding and the observed discrepancy in this study was partly related to lower awareness and treatment rate among patients with IDH in China. On account of the traditional concept that “SBP matters most” in association with cardiovascular events and hypertension mostly asymptomatic , the awaIsolated diastolic hypertension results from an increase in peripheral vascular resistance and is more prevalent in young and middle-aged adults , 38–40. Furthermore, whether IDH needs treatment is controversial and the treatment rate of IDH is low . In our This systematic review and meta-analysis had a number of strengths. First, the meta-analysis included close to 500,000 participants, providing sufficient statistical power to detect associations between IDH and cardiovascular events. Second, this meta-analysis was based on several cohort studies from various populations such as Asian, American, and European, which strengthened the generalizability of the findings. Third, the inclusion of cohort studies ensured that the exposure preceded the outcome, reduced the potential selection bias, and avoided recall bias. Furthermore, all of the included studies were of high quality and sensitivity analyses, further ascertained the robustness of the results.This meta-analysis had several limitations. First, the definition of composite cardiovascular events was somewhat inconsistent in the included studies, which led to some bias; however, we defined the composite cardiovascular events as coronary heart disease, strokes, heart failure, and/or cardiovascular mortality; all data in this study were extracted according to the definition. Second, there was significant between-study heterogeneity as well. When performing subgroup analysis based on the measurement methods of BP for the composite cardiovascular events, the heterogeneity in the mercury sphygmomanometers group and the automatic digital BP monitor group disappeared and the results remained consistent. In consequence, the measurement methods of BP may be the main source of heterogeneity. Mercury sphygmomanometers and automatic digital BP monitors are two main methods of measuring DBP, while the mechanisms differ slightly for DBP . Previous study indicated that the automatic BP device may underestimate DBP by up to 3 mm Hg . FinallyThis meta-analysis provides evidence that IDH defined using the 2018 ESC criterion is significantly associated with an increased risk of composite cardiovascular events, cardiovascular mortality, all strokes, and HS, but not significantly associated with all-cause death and IS. Subgroup analysis further indicates that the correlation between IDH and composite cardiovascular events is significant in younger people and Asians. Therefore, further studies are needed to clarify the age-stratified associations of IDH with cardiovascular events and attach importance to the young population and Asians with IDH. Furthermore, the results of subgroup analysis stratified by treatment status at baseline indicate that active treatment of IDH is helpful in reducing the risk of long-term composite cardiovascular events. However, there is no evidence from clinical trials on the efficacy of antihypertensive medications on BP reduction and long-term cardiovascular events in IDH. In consequence, future studies are needed to assess the impacts and cost-effectiveness of non-pharmacological and pharmacological treatments of IDH for reducing the risk of cardiovascular events.The original contributions presented in the study are included in the article/All authors listed have made a substantial, direct, and intellectual contribution to the work and approved it for publication.This study was supported by the Youth Talent Promotion Project of China Association for Science and Technology (No.2020-QNRCI-02) and the Science and Technology Innovation Project of China Academy of Chinese Medical Sciences (NO.CI2021A05013).The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


Opium is among the most used substance of abuse worldwide. More than 50 million opium users are there worldwide, majority of whom are in Asia. Opium usage have been reported to be associated with cancer. This study aimed to find the association between opium use or abuse and head and neck cancer. st November 2019. Two authors independently reviewed the studies, did quality assessment, and extracted data in standardized data extraction form. Pooled estimate of OR for risk of head and neck cancer was calculated using random effects model using the method of DerSimonian and Laird, with the estimate of heterogeneity being taken from the inverse-variance model. Subgroup and sensitivity analysis were performed. The protocol was registered in PROSPERO (CRD42020156049). A systematic search was conducted in Medline, Scopus, Cochrane, and Google Scholar database for studies published from inception till 1Fourteen studies were included in data synthesis (11 case control studies and 3 cohort studies). Eleven case control studies were included in synthesizing the results for meta-analysis. Pooled odds ratio for risk of cancer among opium users for the 11-case control study was 3.85 (2.18-6.79). Heterogeneity was high . There was no publication bias in the study. Subgroup analysis showed highest OR for pooled estimate for risk of laryngeal cancer (19.98 (11.04-36.15)). There was almost four-fold rise in risk of head and neck cancer among opium users compared to non-users.  Illegal drug use is a worldwide phenomenon, which has drastic effects on public health. Among various drugs of abuse, opiates are among the most used substances and pose a large public health problem around the world. According to world drug report 2019, approximately 53.4 million people globally had used opioids in the year 2017, which is 56 per cent more than the estimate for 2016. ., 2019).Estimates from 2019 suggest that there are approximately 54 million opium users worldwide, 80% of whom live in Asia. Ease of access, and traditional beliefs about opium are the two factors responsible for high rate of consumption  Opium use or abuse leads to several acute health hazards and some reports also states association between use long term effects of opium use and risk of cancer. Link between opium and oral, bladder, lung, head and neck and gastric cancers have been found in past studies.  Many experiments and case control studies have explored the carcinogenic effects of long term use of opium. Opium has a complex chemical composition comprising at least 25 alkaloids and other ingredients. Before consumption as opium, the harvested latex of poppy is dried and boiled. Commonly consumed forms of illicit opium include raw (or crude) opium, opium dross (tarry residues formed after smoking raw opium), and refined opium or opium sap (boiled opium dross with or without added raw opium). All three forms are either smoked or ingested. Smoking any form of opium give rise to opium pyrosylates (solid residues of combusted opium) . Various previous studies have shown carcinogenic characteristics exhibited by opium dross, and opium pyrosylates . Although most of the published studies have not assessed different types, routes and doses of opium which differ in their carcinogenicity, bioavailability and by products .CIA world factbook, states that, India is the world’s largest manufacturer of legal opium for the pharmaceutical industry. Some places where opium are grown are Pratapgarh in Rajasthan; Mandsaur, Ratlam, Neemuch in Madhya Pradesh and Barabanki, Bareily, Lucknow and Faizabad in Uttar Pradesh.  Afghanistan is among the other major cultivator of opium in the world contributing almost half . According to “Global Cancer Incidence, Mortality and Prevalence 2018 Report”, 5 years prevalence for lip and oral cavity cancer in India was 19.59%, with mortality risk of 10.09% whereas head and neck cancer in India accounted for 30% of all cancers .In this systematic review and meta-analysis, we aimed to find the association between opium use or abuse and head and neck cancer. Protocol and Registration: The protocol of this systematic review was registered in PROSPERO  at www.crd.york.ac.uk under the PROSPERO-ID CRD42020156049.The review team developed the methods for the present systematic review and meta-analysis following the PRISMA statement for preferred reporting items for systematic reviews and meta-analysis protocols . The present systematic review and meta-analysis has included published articles if they met all of the following criteria: original observational studies on human participants; exposure related to opium; outcome of interest is head and neck cancer; studies providing risk estimate or enough information to calculate such a measure. Search strategyA thorough systematic literature search was conducted in the following electronic databases namely Medline (PubMed), Cochrane library (Cochrane), Scopus, and Google scholar by two independent researchers. Articles were searched from inception till 1st November 2019. Detailed Search strategy is described in Appendix 1. Rayyan software was used for including, and excluding the articles from the records identified through search strategy. References of included articles were manually retrieved. The whole process of study selection is summarized in PRISMA in Data extraction and managementData was extracted and tabulated using a standardized data extraction form by two independent reviewers. Discrepancies were resolved via referencing the original article. The following data was extracted - author, year of publication, journal, country or region, study design, sample size, diagnostic criteria of head and neck cancer, number of cases and control/ exposed and non-exposed, head and neck cancer type, Opium usage, risk estimate of head and neck cancer comparing various levels of opium usage, and main outcome (Odds ratio). Appraisal of quality and risk bias:The risk of bias was assessed based on the New Castle Ottawa Scale . Eleven Case control studies were analyzed for risk of bias, based on questions asked in following sections: Selection of cases and controls; Comparability of cases and controls; Ascertainment of exposure. Depending on the star rating of each questions of three sections studies were given color coding. If either of two star or both stars present then green, both absent then red and if not mentioned in the study then it is represented by yellow color. Three cohort studies were analyzed based on following questions: Selection of cohorts; Comparability of cohorts; Assessment of outcome . Data synthesis2<25%, heterogeneity was considered as low. Pooled estimate of OR for risk of head and neck cancer was calculated using random effects model using the method of DerSimonian and Laird, with the estimate of heterogeneity being taken from the inverse-variance model. Forrest plot was made for pooled OR of included case control studies. Subgroup and sensitivity analysis were performed. All analyses were carried out in STATA software .The risk estimate (OR) of head and neck cancer due to opium usage was provided with 95% confidence interval (C.I.). Heterogeneity across enrolled studies was quantified using the Q-statistic and inconsistency index (I2). I2>50%, heterogeneity was considered as severe; if I2=25% to 50%, heterogeneity was considered as moderate, and if IPublication Bias - Publication bias was analyzed and represented by a funnel plot, and funnel plot symmetry was assessed with Egger’s test . STATA sTotal 324 studies were identified through various databases . After removal of duplicates, 262 studies were screened out of which 159 full text articles were assessed for eligibility and based on exclusion and inclusion criteria 145 studies were excluded. Out of the 145 studies which were excluded, in 101 studies risk of opium with cancer was not studied and in remaining 44 studies, risk of opium was studied with cancer other than head and neck cancer. 14 studies were included in data synthesis (11 case control studies and 3 cohort studies). Details of the studies included are given in Shakeri et al.,(2012), Nasrollahzadeh et al., (2008), Islami et al., (2004), and Pournaghi et al., (2019) studied the relationship between opium consumption and esophageal squamous cell carcinoma with odds ratio (OR) ranging from 1.37 to 2.1, whereas for oral cancer OR ranged from 4.09 to 5.01 as described in studies done by Fahmi et al., (1983), Razmpa et al.,(2014), and Saedi et al.,(2012). The risk of development of laryngeal cancer with opium consumption was highest and the OR ranged from 9.09 to 31.55 according to the studies done by Mausavi et al.,(2010), Berjis et al., (2017), and Bakhshae et al., (2017). There were three cohort studies namely by Aghcheli et al., (2011), Sheikh et al., (2019), and Khademi et al.,(2012) which reported hazard ratio ranging from 1.23 to 1.98 among opium users for the risk of esophageal squamous cell carcinoma.Quality assessment of studies was done using New Castle Ottawa Scale. Details of risk of bias assessment is shown in detail in 11 case control studies were included in synthesizing the results for meta-analysis. The pooled odds ratio for risk of cancer among opium users for the 11-case control study was 3.85 (2.18-6.79). Heterogeneity in the pooled estimate was high .  In threePublication bias: Sub-group analysisThe high heterogeneity in the pooled estimate was further investigated through subgroup analysis. Highest OR was observed for the pooled estimate for risk of laryngeal cancer (19.98 (11.04-36.15)). Odds ratio for risk for esophageal cancer was observed as 1.66 (1.37-2.02), OR for risk of oral cavity cancer was observed as 3.06 (1.29-7.29). Sensitivity analysisWe have included only head and neck cancer for our systematic review and meta-analysis, but the study by Tahami et al., (2014) and Islami et al., (2004) included other upper GI cancers as well, therefore we performed sensitivity analysis dropping the study by Tahami et al., (2014) and Islami et al., (2004) which did not cause significant change in the pooled estimate. ; compared to before 3.85 (2.18-6.79)). We also did sensitivity analysis by dropping small studies (sample size less than 200); however, the pooled estimate did not change after that also. ; compared to before 3.85 (2.18-6.79)).Opioids, which include opiates (heroin and opium) and pharmaceutical and other synthetic opioids, are a major concern in many countries because of the severe health consequences associated with their use. In 2017, the use of opioids accounted for nearly 80 per cent of the 42 million years of “healthy” life lost as a result of disability and premature death  and 66 per cent of the estimated 167,000 deaths attributed to drug use disorders. Asia, host to more than 90 per cent of global illicit opium production and the world’s largest consumption market for opiates, accounted for almost 80 per cent of all opiates seized worldwide Middle East and South-West Asia and South Asia are two subregions that together account for almost 60 per cent of the estimated number of opiate users worldwide. The largest quantities of both opium and morphine seized were reported by the Islamic Republic of Iran, followed by Afghanistan and Pakistan. Iran is one of the topmost countries where largest quantities of opiates were seized.A major drug use survey carried out recently in India found that in 2018, 2.1 percent of the population aged 10–75, a total of 23 million people, had used opioids in the past year. Among opioids, heroin is the most prevalent substance, with a past this is followed by the non-medical use of pharmaceutical opioids, and by opium at almost 0.5 per cent. Of the 23 million past-year opioid users, roughly one third, or 7.7 million people, suffer from opioid use disorders. Estimated levels of drug use disorders were: 0.1 per cent for opium use; 0.57 per cent for heroin use; and 0.23 per cent for use of pharmaceutical opioids. Compared with earlier estimates from a survey carried out in 2004, overall opioid use in India is estimated to have increased fivefold. Various mechanism has been proposed for the link between opium and different cancers . The first potential mechanism is the genotoxic or mutagenic effect of opium smoke and pyrolysates, and some opium Alkaloids. During opium pyrolysis, multiple carcinogenic compounds are produced including heterocyclic and polycyclic aromatic hydrocarbons, primary aromatic amines, and N-nitrosamines, which can enter the body through the respiratory and digestive tracts and affect different organs .Pyrolyzed derivatives and other mutagens containing hydroxy-phenanthrene moiety have been found in opium which have genotoxic and carcinogenic properties. . Sister chromatid exchange, and frame shift mutation have been observed in human peripheral blood lymphocytes and Salmonella typhimurium due to pyrolysis derived nitrogen containing heterocyclic components opium . Morphine has also been found to be linked with carcinogenesis, by increasing methylation of DNA . The second potential mechanism is through the tumor promoting effects of opiates  Opium has also been found to be associated with other aspects of carcinogenesis, like angiogenesis, epithelial-mesenchymal transition, and proliferation in animal models and human tissues .The third potential mechanism is through facilitation of the effects of other carcinogens on different tissues, either by modifying the pharmacokinetics of these carcinogens and increasing their bioavailability or by impairing the physiological function of some organs and thus prolonging their exposure to the potential carcinogens .Other mechanism hypothesized as for carcinogenesis due to opium use is adulteration of opium with heavy metals during its preparation .These studies showed causal association between opium and cancer. Previous observational studies have shown association between opium addiction and higher risk of various types of cancers, namely pancreas , oral cavity , esophagus , stomach , lung , larynx , and bladder . Most of the studies showed different estimates for opium-only consumption, opium plus cigarette use, and opium use regardless of smoking status. Different types of opium, their methods of ingestions, duration of opium use, and the dose of opium, is associated with various cancers. A systematic review done by Afshari et al., (2017). Iran for opium use and risk of bladder cancer depicted the OR of 3.85 (3.05-4.87). Present meta-analysis reported higher OR for risk associated with development of head and neck cancer in opium users compared to non-users.In another study done by Sheikh et al., (2019) Opium users have a significantly higher risk of developing cancers in different organs of the respiratory, digestive, and urinary systems and the CNS and that regular use of opiates might increase the risk of a range of cancer types. Use of opium was associated with an increased risk of developing all cancers combined , gastrointestinal cancers , and respiratory cancers  in a dose-dependent manner. For site-specific cancers, use of opium was associated with an increased risk of developing esophageal , gastric , lung , bladder , and laryngeal  cancers in a dose-dependent manner. High-dose opium use was associated with pancreatic cancer . Ingestion of opium (but not smoking opium) was associated with brain  and liver  cancers in a dose-dependent manner.In present study the pooled estimates from the eleven case control studies included shows approximately four-fold increase in the odds of developing head and neck cancer with the use of opium. This is the first systematic review and meta-analysis in the world, conducted on risk of head and neck cancer due to opium usage. The review provided an opportunity to examine the influence of opium on the risk of head and neck cancer. There was no evidence publication bias, which further increased the confidence in our estimate. There were certain limitations with the study mentioned which are given below. The heterogeneity was high which could be due to different methodologies, and different criteria for the exposure variable followed among studies included in the meta-analysis. Matching the case-control groups as well as the confounders adjusted during the odds ratio estimations were not same among the studies. Effect of opium should be estimated independently from tobacco smoking, which is a proven risk factor for many cancers. The results reported by the case-control studies might be prone to biases. It is better to investigate the associations between opium use and cancer by conducting longitudinal studies. We included three longitudinal studies in our current meta-analysis and in one of the cohort studies, the hazard ratio for risk of esophageal cancer among opium users was 1.85. Also, all the studies included were conducted in Iran. In the current metanalysis, we investigated did not consider the effect of the opium consumption methods. It should be taken into consideration by the future studies. The results of this meta-analysis can provide important information for the general population to protect themselves against a proven risk factor for one of the highly virulent cancers. People should not only prevent and stop smoking habits, but also avoid other behavioral risk factors such as opium consumption. In conclusion, this study showed that there was almost four-fold rise in risk of head and neck cancer among opium users compared to non-users. Although, there was large heterogeneity in the pooled estimate which could be due to difference in measure of opium usage. Also, the between-study variance was high which can be due to different type of head and neck cancers in different studies. Finally, Considering the increasing trend of opium consumption among populations especially in countries such as Iran which is in a critical gateway of opium substances, similar studies need to be carried out in India which is the largest manufacturer of legal opium for the pharmaceutical industry.AJ and GS conceived and designed the study; AJ, GS, and ADG acquired data; AJ, GS, and ADG analyzed, and interpreted data; GS, and ADG drafted the initial manuscript. ADG, PRR, GS, and AJ performed critical revisions of the manuscript, and finalized the manuscript. All the authors approved the final version of the manuscript.The protocol of this study was registered in PROSPERO  at www.crd.york.ac.uk under the PROSPERO-ID CRD42020156049.


Multifarious factors have a causal relationship with gastric cancer (GC) development. We conducted a comprehensive analysis to evaluate the strength of the evidence examining non-genetic risk factors for gastric cancer.PubMed, Web of Science, and the Cochrane Library were searched from inception to November 10, 2021 to identify meta-analyses of observational studies examining the association between environmental factors and GC risk. For each meta-analysis, the random effect size, 95% confidence interval, heterogeneity among studies, and evidence of publication bias were assessed; moreover, the evidence was graded using predefined criteria, and the methodological quality was evaluated using AMSTAR 2.p < 0.05). Only 10 associations had strong epidemiologic evidence, including 2 risk factors (waist circumference and bacon), and 8 protective factors ; 26 associations had moderate quality of evidence; and the remaining 57 associations were rated as weak. Ninety-four (68.61%) associations showed significant heterogeneity. Twenty-five (18.25%) associations demonstrated publication bias.A total of 137 associations were examined in 76 articles. Among these meta-analyses, 93 associations yielded significant estimates (In this comprehensive analysis, multiple associations were found between environmental factors and GC with varying levels of evidence. Healthy dietary habits and lifestyle patterns could reduce the risk for GC. However, further high-quality prospective studies are still necessary to draw more definitive conclusions. The incidence of gastric cancer (GC) has gradually decreased in recent decades, mainly due to improved socioeconomic status, hygienic practices, and consequentially reduced Helicobacter pylori (HP) infection rates –3. HowevAlthough many systematic reviews and meta-analyses have examined risk factors for GC , to our This study was registered in the International Prospective Register of Systematic Reviews (registration number: CRD42021290515). This study did not require ethical approval.Two observers (HYQ and TCY) independently and systematically searched PubMed, Web of Science, Cochrane Library from inception to November 10, 2021, to identify observational studies of systematic reviews and meta-analyses assessing the association between non-genetic risk factors and GC using the following search algorithm:  AND . The reference lists of retrieved eligible papers were further hand-searched to avoid missing other potentially related studies. Only articles published in English were included.Articles were deemed qualified if they satisfied all of the following inclusion criteria: (1) the articles were systematic reviews and meta-analyses of observational studies ; (2) the study evaluated the association of environmental (non-genetic) factors and GC risk, but not for screening, diagnostic, prognostic purposes; and (3) the study provided enough data to perform the analyses. The title and abstract of all eligible papers were initially screened, and then the full text of possible qualified articles was retrieved for further perusal based on the pre-established inclusion and exclusion criteria. Disagreements between two investigators were settled through a discussion. If multiple meta-analysis examined the identical scientific issue, we chose the largest number of studies to avoid repeated evaluation of the same primary studies , 11.From each included meta-analysis, two researchers (HYQ and TCY) independently extracted the following data: the examined risk factors, the first author's name, year of publication, the epidemiological design and number of included studies, and the number of participants and cases. The study-specific relative risk estimates , heterogeneity, and publication bias for every risk factor were also collected in each study. Divergence during data extraction was clarified by discussion until a consensus was reached.Two researchers (HYQ and TCY) independently appraised the methodological quality of all included systematic reviews using the updated 16-item AMSTAR 2 instrument  . This toTwo authors (HYQ and TCY) independently assessed the strength of the epidemiologic evidence using the following criteria , 13, 14:p-value <0.001 , and . The strength of the epidemiologic evidence was classified as high , moderate , or weak . When the p-value was not directly reported, it was recalculated from the 95% confidence interval of the pooled effect estimate using an established method ((1) precision of the estimate , and  no evided method . In casep-value of the pooled estimate of effect size of < 0.05 was deemed significant. I2 and Q test was used to determine the heterogeneity  indicated a significant heterogeneity and a publication bias. Values (I2 test) exceeding 50% were generally considered to indicate high heterogeneity. All p-values were two tailed, and all statistical analyses were performed using Stata version 16.0.Based on the extracted raw data from every included study, we reanalyzed and presented the random-effects estimate whenever the fixed-effects model was initially used . And in ogeneity  among stogeneity ; a p-valp-value of <0.05, while 47 (34.3%) reported a p-value of <0.001 , of which 76 were finally deemed eligible –95. The f <0.001 .Obesity is a well-recognized risk factor for multiple adverse health outcomes. Waist circumference and waist–to–hip ratio were associated with an increased risk of GC in the highest vs. lowest comparisons  . SimilarThe Mediterranean diet (MedDiet) and dietary total antioxidant capacity (D-TAC) were associated with significant reductions in GC risk for the highest vs. lowest comparisons  , 26. FibHigher vitamin consumption was associated with reduced GC risk  , especiaRegular proton pump inhibitors (PPI) use was associated with GC  . ConversSmoking can be linked to the development of many cancers, including GC , 74. PhyDepression was associated with an increased GC risk  . Non-alcIn addition to well-recognized HP and Epstein-Barr virus, other viruses  were also associated with a higher incidence of GC –89 Figu.A strong inverse association was found between socioeconomic position indicators  and GC risk  . Of note2 < 50% and p-value of Cochran Q test ≥ 0.10), while the remaining 94 (68.61%) associations indicated significant heterogeneity . In terms of publication bias, 110 (80.29%) meta-analyses showed no evidence of significant small-study effects , whereas 25 (18.25%) meta-analyses demonstrated publication bias . With regard to the associations of GC with fiber intake (10 g/d increment)  associations showed absence of heterogeneity (Icrement)  and HTLVcrement) , small-sn = 64, 84.21%), incomplete literature search , and the absence of list of excluded studies .The methodological qualities of the 76 included articles were assessed and graded using the 16-item AMSTAR2 tool ; three . Based op-value of < 0.001, and absence of large heterogeneity and small-study effects), including two risk factors (waist circumference and bacon) and eight protective factors  (The outcomes of the epidemiologic evidence measurement are presented in rtility) , 72, 94.rtility) .In this comprehensive analysis, 76 meta-analyses of observational studies were identified and appraisal of current evidence that examined the association of GC with various environmental risk factors was performed. All 137 environmental associations, covering anthropometric indices, dietary intake, micronutrients, use of medication, lifestyle, pre-existing medical history, viral or bacterial infection, and others, were assessed. Among these, two risk factors (waist circumference and bacon) and eight protective factors  yielded high epidemiologic evidence without any evidence of bias , 72, 94.The association between waist circumference and increased risk of GC was supported by high epidemiologic evidence , indicatOf note, more than half of the associations examined a broad variety of dietary factors, which revealed the current direction of this line of research. The MedDiet is a recognized healthy dietary pattern, characterized by relatively high consumption of fruits, vegetables, legumes, and olive oil, moderate intake of dairy products and fish, and very limited intake of red meat and processed meat products , 102. OuWith regard to medication therapy, long-term use of PPIs could increase GC risk, as supported by moderate epidemiologic evidence, mainly due to the accelerating progression of HP-related atrophic gastritis- and hypergastrinemia-promoting enterochromaffin-like cell hyperplasia . HoweverThe estimated 60 well-established carcinogens found in cigarette smoke could explain the positive association between smoking and GC risk . ModeratMost of the assessed associations could not show high epidemiological evidence due to the significant heterogeneity and/or small-study effects. Heterogeneity usually indicates the presence of bias in some primary studies, but might also be due to the real differences among studies. Genuine heterogeneity might play a role in the field of GC, including the difference in exposure assessment, the mixture of cohort and case-control studies in some meta-analyses, differential association of risk factors due to geographical heterogeneity, and so on. As positive results are more likely to be published compared with null results, and the study participants may be a small portion of the actual population with the disease, the probability of small-study effects should be taken into consideration. The reported associations with GC need to be explained with caution, particularly in meta-analyses with a relatively small number of included studies; the heterogeneity and publication bias among researches are evident.Our study has several strengths. This comprehensive analysis was the first to provide a comprehensive overview of the evidence to evaluate the association of non-genetic factors with GC risk, although several studies reported the risk factors of GC. Then, the systematic literature search, article selection, and data extraction were conducted by two independent researchers. Additionally, the AMSTAR 2 tool was utilized to appraise the methodological quality of the eligible systematic reviews. Furthermore, the epidemiologic evidence was graded in accordance with the predefined criteria including the evaluation of the estimate precision, heterogeneity, and publication bias.Nevertheless, our study has several limitations. First, only published meta-analyses of observational researches were included; thus, other risk factors with enough evidence that have not yet been evaluated by meta-analytic quantitative synthesis were possibly overlooked. Second, the quality of component meta-analyses was not evaluated as it exceeded the range reported in our study, and meta-analyses should be conducted by the researchers of the primary studies. Third, the majority of included meta-analyses showed heterogeneity, and the observational investigations were prone to selection and recall biases, especially case-control studies. Fourth, the systematic reviews and meta-analyses contained in our study were only published in English; thus, the possible missing information that were published in other languages might affect the evaluation results.In conclusion, developing a healthier dietary  and lifestyle pattern along with promoting physical activity to prevent obesity could hopefully reduce the incidence of GC in the near future. However, further high-quality prospective studies excluding potential residual confounders are needed; the application of reporting guidelines   and regiThe original contributions presented in the study are included in the article/YH and XG contributed to the concept and design of the comprehensive analysis. YH, CT, and DL collected and analyzed the data. YH and HZ drafted the manuscript. XG revised the manuscript. All authors read and approved the final manuscript to be published.


In spite of substantial declines in both incidence and mortality rates in the past 50 years, cervical cancer remains one of the leading causes of cancer associated mortality among women globally. We performed this meta-analysis to explore the role of XRCC3 rs861539, MTHFR rs1801133, IL-6 rs1800795, IL-12B rs3212227, TNF-α rs1800629 and TLR9 rs352140 polymorphism with susceptibility to cervical carcinoma. The search databases include PubMed, SciELO, MedRxiv, Web of Science, Scopus, Cochrane Library, China National Knowledge Infrastructure, and China Biology Medicine disc up to 30 June 2021. The language is limited to English and Chinese. The comparison between the polymorphisms and cervical cancer was assessed using pooled odds ratio (OR) and 95% confidence interval (CI). The data are statistically analyzed by Comprehensive Meta-Analysis (CMA) 2.0 software. A total of 59 studies including seven studies with 1,112 cases and 1,233 controls on XRCC3 rs861539, 14 studies with 2,694 cases and 3349 controls MTHFR rs1801133, four studies with 1,121 cases and 1,109 controls on IL-12B rs3212227, seven studies with 1,452 cases and 2,186 controls on IL-6 rs1800795, 20 studies with 4,781 cases and 4909 controls on TNF-α rs1800629, and seven studies with 1743 cases and 2292 controls on TLR9 rs352140 were included. There was a significant association between XRCC3 RS861539, TNF-α rs1800629, and IL-6 rs1800795 polymorphisms and an increased risk of cervical carcinoma in overall population. However, the MTHFR rs1801133, IL-12B rs3212227 and TLR9 rs352140 polymorphisms were not associated. The pooled analysis showed that XRCC3 RS861539, TNF-α rs1800629, and IL-6 rs1800795 were associated with cervical carcinoma susceptibility, but not MTHFR rs1801133, IL-12B rs3212227 and TLR9 rs352140 polymorphisms. Cancer is one of the main public health problems globally with about 18.1 million new cancer cases and 9.6 million cancer deaths in 2018 . The World Health Organization (WHO) states cervical cancer is the second most frequent cancer among women worldwide  with an estimated 570,000 new cases and 311,000 deaths in 2018 worldwide . It is reported that 85% of cervical cancer cases and 87% of the cervical cancer deaths occur in less developed countries . With over 500,000 cases of cervical cancer reported each year, nearly 80 percent of those are in developing countries, including Africa with 68,000, an estimated 77,000 in Latin America and the Caribbean, and 245,000 in Asia . Cervical cancer is driven by persistent infection with one of 15 carcinogenic human papillomavirus (HPV) types . However, the impact of HPV type and intratypic variants on patient outcomes is far less understood .To date, the occurrence and development of cervical cancer is suggested to be associated with persistent HPV infection . The DNA of HPV integrates into the host cell genome and disrupts the open reading frame and causes overexpression of E6 and E7 genes . However, the specific molecular mechanisms and potential single gene require further studies. Familial based studies and evaluation of inherited genetic variations revealed that host genetic factors have a role in cervical cancer pathogenesis . Recently, a comprehensive review study showed that CDK1, CCNB1, ITGB1, FN1, MMP9 and STAT1 played different roles in the progression of cervical cancer through different signaling pathways . Moreover, studies have shown that cervical cancer has a heritable genetic component. However, as other complex disease, the identified loci only explain a minority of the risk of cervical cancer. This gap of the heritability explained by genetic markers genome-wide association studies and the heritability identified in familial studies has been termed as ‘missing heritability’ . Thus, our understanding of the genetic basis of cervical cancer is still limited. In this meta-analysis, we explore the role of XRCC3 rs861539, MTHFR rs1801133, IL-6 rs1800795, IL-12B rs3212227, TNF-α rs1800629, and TLR9 rs352140 polymorphisms in susceptibility to cervical cancer.Literate Search StrategyWe performed a comprehensive literature search on PubMed, Scopus, China National Knowledge Infrastructure, Wanfang, VIP Information Chinese Journal Service Platform, and China Biology Medicine disc, PubMed, EMBASE, Web of Science and the Cochrane Library, Google Scholar, Cochrane Library, EMBASE, Scientific Information Database (SID), WanFang, VIP, Chinese Biomedical Database (CBD), Scientific Electronic Library Online (SciELO), China National Knowledge Infrastructure (CNKI), IranDoc and Egyptian Knowledge Bank (EKB) Journals database to identify all relevant studies on the association of XRCC3 rs861539, MTHFR rs1801133, IL-6 rs1800795, IL-12B rs3212227, TNF-α rs1800629 and TLR9 rs352140 with cervical cancer up to 30th June 2021. The following terms were used in various combinations in this search:  AND (‘’X-Ray Repair Cross Complementing 3’’ OR ‘’XRCC3’’ OR ‘’rs861539’’ OR ‘’g.4880C>G’’ OR ‘’c.-237C>G’’ OR ‘’n.54-321G>C’’) AND  AND (‘’Interleukin 12’’ OR ‘’IL-12’’ OR ‘’1188A>C’’ OR ‘’rs3212227’’ OR ‘’ g.158742950T>G’’ OR ‘’ g.19532A>C’’ OR ‘’ c.*159A>C’’) AND (‘Interleukin 6’’ OR ‘’IL-6’’ OR ‘’-174G>C’’ OR ‘’rs1800795’’ OR ‘’ g.4880C>G’’ OR ‘’c.-274C>G’’ OR ‘’ n.54-321G>C’’) AND (‘’Toll like receptors’’ OR ‘’TLR’’ ‘’ OR ‘’rs352140’’ OR ‘’2848G/A’’) AND (‘’Gene’’ OR ‘’Single-Nucleotide Polymorphism” or “SNP” OR ‘’Polymorphism’’ OR ‘’Genotype’ OR ‘’Allele’’ OR ‘’Variant’’ OR ‘’Variation’’ OR ‘’Mutation’’ OR ‘’Mutant’’). The search was carried out in English, Chinese and Persian. When overlapping data on the same cases were included in more than one publication, only the one with the larger sample size was selected. Moreover, the reference list of the retrieved studies and reviews manually checked to identify more potential eligible studies.Eligibility criteriaTo obtain the papers, the following criteria should be met to include the papers in our study: 1) studies with case-control or cohort design; 2) studies reported original data; 3) studies appraised association of XRCC3 rs861539, MTHFR rs1801133, IL-6 rs1800795, IL-12B rs3212227, TNF-α rs1800629 and TLR9 rs352140 polymorphisms with cervical cancer; and 4) studies with available and sufficient data for calculating an odds ratio (OR) with 95% confidence interval (CI). The following exclusion criteria were also used: 1) studies on other polymorphism at XRCC3, MTHFR, IL-12B, IL-6, and TNF-α genes; 2) animal studies or in vitro studies; 3) studies with in sufficient data on genotype frequencies or which the number of genotypes and alleles could not be ascertained; 4) linkage studies; 5) family based studies including sibling, twins and trios-parents studies; 6) abstracts, case reports, commentaries, editorials, conference articles, reviews, proceedings and meta-analyses; and 7) duplicates studies or overlapping data.Data ExtractionTwo authors reviewed and extracted necessary information independently in accordance with our inclusion criteria. For conflicting data, the authors carried out discussions until a consensus was reached. If they could not reach a consensus, disagreement was resolved by the third author who participated in the discussion. For each eligible study the following data was collected: first author name, year of publication, country of origin, ethnicity , genotyping methods, sample size, allele and genotype frequency of XRCC3 rs861539, MTHFR rs1801133, IL-12B rs3212227, IL-6 rs1800795, TNF-α rs1800629 and TLR9 rs352140 polymorphisms in cervical cancer cases and controls, Minor Allele Frequency (MAFs) and Hardy-Weinberg equilibrium (HWE) in healthy controls. In this meta-analysis different case-control groups or cohorts in one publication were considered as independent studies. We did not define any minimum sample size to include in this meta-analysis. The ‘‘mixed’’ group means mixed or unknown populations. If more than one study was published by the same author(s) using repeated or overlapped data, the studies with the largest sample size or the most recently published study was included to the meta-analysis. If selected articles did not reported necessary data the corresponding authors was contacted by email to request the missing data.Statistical Analysisχ2 test and I2 statistics were used to assess whether there was between-study heterogeneity, in which P >.10 and I2 < 50% could be considered that there was no statistical heterogeneity between the research results. The fixed effect model was selected for data consolidation; P ≤.10 and I2 ≥50% could be considered that there was statistical heterogeneity between the research results, and a random effect model was used for data consolidation. If there was significant heterogeneity among studies, the random effects model (DerSimonian and Laird) was used; otherwise, the fixed-effects model (Mantel and Haenszel) was acceptable. The Hardy-Weinberg equilibrium of the control group was evaluated using the χ2 test, and the expected and actual genotype frequencies of the control group were compared. In this meta-analysis, P-values of ≤.05 were considered statistically significant . In addition, the sensitivity analysis was performed by excluding HWE-violating studies. Potential publication bias was evaluated using the Egger’s linear regression test and Begg’s quantitative test. The asymmetric plot of Egger’s test and the P-value of Begg’s test less than 0.05 were considered a significant publication bias. If there was evidence of publication bias (P<0.05), trim and fill method was applied to adjust for the effect of publication bias . All statistical analyses were performed using Comprehensive Meta-Analysis (CMA) Software version 2.0 . All tests were two-sided, and the P< 0.05 was considered statistically significant.The comparison between the XRCC3 rs861539, MTHFR rs1801133, IL-6 rs1800795, IL-12B rs3212227, TNF-α rs1800629 and TLR9 rs352140 polymorphism and cervical cancer was assessed using pooled odds ratio (OR) and 95% confidence interval (CI). The pooled data were calculated under five genetic models, i.e., allele (B vs. A), homozygote (BB vs. AA), heterozygote (BA vs. BB), dominant (BB+BA vs. AA) and recessive (BB vs. BA+AA), in which a ‘‘A’’ denotes a major allele; ‘‘B’’ denotes a minor allele. The Characteristics of the Enrolled StudiesThe flowchart of selection of studies and reasons for exclusion is presented in Figure 1. There were 1,013 articles relevant to our search words and 395 duplicated studies were excluded. Then, 207 studies removed after reading titles and abstracts. Following the inclusion exclusion criteria 352 studies were excluded. Finally, a total of 59 studies were included in this meta-analysis. There are seven studies with 1112 cases and 1233 controls on XRCC3 rs861539 , 14 studies with 2694 cases and 3349 controls MTHFR rs1801133 , four studies with 1121 cases and 1109 controls on IL-12B rs3212227 , seven studies with 1452 cases and 2186 controls on IL-6 rs1800795 , 20 studies with 4,781 cases and 4909 controls on TNF-α rs1800629  and seven studies with 1743 cases and 2292 controls on TLR9 rs352140 . The details of included studies were shown in Quantitative Data SynthesisXRCC3 RS861539The summary for the association of XRCC3 RS861539 polymorphism with cervical cancer risk are shown in TNF-α rs1800629The summary for the association of TNF-α rs1800629 polymorphism with cervical cancer risk are shown in MTHFR rs1801133The summary for the association of MTHFR rs1801133 polymorphism with cervical cancer risk are shown in IL-12B rs3212227The summary for the association of IL-12B rs3212227 polymorphism with cervical cancer risk are shown in IL-6 rs1800795The summary for the association of IL-6 rs1800795 polymorphism with cervical cancer risk are shown in TLR9 rs352140The summary for the association of TLR9 rs352140 polymorphism with cervical cancer risk are shown in Test of heterogeneityχ2 test and I2 statistics were used for assessing the heterogeneity of the included studies. Results indicate that there was statistical heterogeneity for XRCC3 rs861539, MTHFR rs1801133, IL-6 rs1800795, IL-12B rs3212227, TNF-α rs1800629, and TLR9 rs352140 polymorphisms under most genetic models. Thus, the random effect model was used for evaluating the pooled OR and 95% CI for those models. Subgroup analyses showed that ethnicity of participants might contribute to part of heterogeneity.In the current meta-analysis the Sensitivity AnalysisThe process of performing a meta-analysis involves a sequence of decisions and it is important to perform a sensitivity analysis in order to assess the impact of different decisions on pooled data. Thus, a sensitivity analysis was carried out by excluding a single study in turn on pooled ORs. The results showed that no individual study had an influence on the pooled OR all involved polymorphisms at XRCC3 rs861539, MTHFR rs1801133, IL-6 rs1800795, IL-12B rs3212227, TNF-α rs1800629, and TLR9 rs352140 polymorphisms, suggesting the stability of our findings. Moreover, excluding HWE deviated studies suggested that there were no independent studies that significantly influenced our pooled data.Publication BiasBegg’s funnel plot and Egger’s test were used for evaluating publication bias. As shown in Genetic factors have been shown to influence the susceptibility of patients to various diseases and have attracted increasing attention . Several efforts have been made to identify the genetic susceptibility factors underlying development of cervical cancer. However, only a few polymorphisms have shown consistency among studies. In this meta-analysis, the association of XRCC3 rs861539, MTHFR rs1801133, IL-6 rs1800795, IL-12B rs3212227, TNF-α rs1800629 and TLR9 rs352140 polymorphisms with susceptibility to cervical carcinoma was assessed by including all relevant studies.Yuan et al., (2021) in a meta-analysis based on 15 studies with 5,740 cases and 9,931 controls revealed that there was no significant association between the XRCC3 Thr241Met and the risk of gynecological malignancies. However, their subgroup analysis by ethnicity showed that XRCC3 Thr241Met was associated with an increased risk of gynecological malignancies in Asians. Moreover, their stratified analysis by cancer type indicated that XRCC3 Thr241Met associated with cervical cancer in Asians . Abbas et al., (2010) in a case-control study with 260 cervical cancer cases and 265 controls evaluated the association of XRCC1+399A/G, XRCC2+31467G/A and XRCC3+18067C/T polymorphisms with cervical cancer in Indian women. Their results showed that XRCC2+31479G/A and XRCC3+18067C/T polymorphisms were not associated with cervical cancer. However, they showed that he XRCC1+399A/G is linked with cervical cancer in the Indian population. Al-Harbi et al., (2017) in a study among 232 cervical cancer cases and 313 control subjects evaluated the association of CDKN1A C31A, ATM G1853A, HDM2 T309G, TGFB1 T10C, XRCC1 G399A, and XRCC3 C241T with cervical cancer among Saudi Arabian women. They showed that the TGFB1 T10C and XRCC1 G399A polymorphisms were associated with cervical cancer risk.Our pooled results showed that the MTHFR rs1801133 polymorphism was not associated with cervical cancer in overall population and by ethnicity. Silva et al., (2019)reported that there were no differences in the genotypic and allelic distribution of MTHFR C677T polymorphism between remission (with the presence of pre-neoplastic lesions) and Persistence (with the presence of pre-neoplastic lesions). The same authors, in another study showed that MTHFR C677T polymorphism was not associated with cervical cancer and HPV infection . However, Sohrabi et al., in a case-control study evaluated the association of MTHFR A1298C and C677T variants among in 50 cervical intraepithelial neoplasia cases, 98 HPV-positive subjects and 47 non-cancerous/non-HPV patients as healthy controls. Their results showed that MTHFR 1298 CC is more likely to be a potential risk factor for HPV-cervical cancer progression . Zhou et al., (2020) indicated that the MTHFR rs4846048 enhanced the risk of cervical cancer through association with miR-522. Gong et al., (2018) reported that MTHFR C677T polymorphism was not associated with the risk of cervical cancer or cervical intra-epithelial neoplasia, while, the MTHFR A1298C polymorphism could increase the risk of both cervical cancer and cervical intra-epithelial neoplasia.In the current study pooled data indicated that the TNF-α rs1800629 polymorphism was associated with cervical cancer risk. Wang et al., (2012) in meta-analysis based 27 studies showed that both TNF-α -238 and -308 G/A polymorphisms could be used to identity individual with elevated susceptibility to cervical cancer in by ethnicity . Behboodi et al., (2021) in a study among 153 Iranian cervical cancer cases and 292 free cancer subjects demonstrated that TNF-α rs1800629 was associated with increased level and risk of developing cervical cancer. However, Duvlis et al., (2020) reported that TNF-a-238G/A and TNF-a-308 G/T polymorphisms were not associated with the risk of HPV associated cervical intraepithelial lesions or cervical cancer cases in Macedonian women compared to controls. Moreover, Traore et al., (2020) revealed that TNF-308 G/A or IL18-607C/A polymorphisms were not associated with HPV infection among Burkina Faso women.Pooled data from all eligible studies showed that the IL-6 rs1800795 polymorphism was significantly associated with cervical cancer risk in overall population and among Asian women. Similarly, Duan et al., (2018) in a meta-analysis based on 7 studies showed that the IL-6 rs1800795 polymorphism is associated with risk of cervical cancer in overall. Pu et al., (2016) in a study with 360 cervical cancer cases and 728 healthy subjects showed that this polymorphism is risk factor for cervical cancer development in Chinese women. de Souza et al., (2006) in a case-control study with 56 cases and 253 controls evaluated the association of IL-6 rs1800795 polymorphism with cervical cancer in a Brazilian population. In 2017, Zidi et al., (2017) in a case-control study evaluated the effects of six different genetic variants atIL-6 with risk of cervical cancer. The study revealed that the IL-6 rs1800795 polymorphism has a has a protective role in cervical cancer development in Tunisian women.de Lima Júnior et al., (2016) revealed that the IL-6 rs1800795 polymorphism was not associated with HPV infection and healthy controls in in Brazilian women. The current meta-analysis data showed that the IL-12B rs3212227 genetic variant was not associated with susceptibility to cervical cancer, while; stratified analysis showed that this polymorphism was associated with cervical cancer risk among Asian women. Zheng et al., (2017) in a meta-analysis based on 33 articles with 10,587 cancer cases and 12,040 healthy subjects assessed the genetic association of IL-12B rs3212227 with cancer risk. Their pooled data indicated that IL-12B rs3212227 polymorphism was associated with cancer risk in overall. However, subgroup analysis by cancer type showed that IL-12B rs3212227 was not associated with cervical cancer. In 2016, Chang et al., (2015) in a meta-analysis based on 5 studies with 2552 cervical cases and 2232 healthy subjects showed that this polymorphism did not associate with risk of cervical cancer. In 2012, de Carvalho et al., found that the IL-12B rs3212227 variant has a protective role in development of cervical cancer in Brazilian women . Similarly, Han et al., (2008) found that IL-12B rs3212227 did not associate with cervical cancer risk in Korean population. However, Roszak et al., (2012) showed that this polymorphism was associated with increased risk of cervical cancer among polish women.The current meta-analysis data showed that TLR9 rs352140 was not associated with susceptibility to cervical cancer. Nath et al., (2020) in a study showed that TLR4/9 polymorphisms are associated with increased HPV16/18 infection susceptibility and cervical squamous cell carcinoma risk among the women of Jharkhand state. In another study in India showed the TLR4 and TLR9 polymorphisms and haplotypes with hrHPV infection and cervical cancer risk . Martínez-Campos et al., (2017) in a case-control study revealed that TLR9 rs187084 is a risk factor for HPV infection, squamous intraepithelial cervical lesion, and uterine cervical neoplasm in Mexican female population. Jin et al., (2017) also evaluated the association of some variants at Toll-like receptors gene with cervical cancer among 420 Chinese cervical cancer patients and 842 controls. Their results showed that mutant alleles of TLR2 rs3775290, TLR4 rs7873784, and TLR9 rs352140 were associated with increased cervical cancer risk. Similarly, Yang et al., (2020) in meta-analysis based on eleven studies indicated that the TLR9 rs187084 and rs352140 polymorphisms may contribute to development of cervical cancer, but not TLR2-196 to -174 del/ins polymorphism.In summary, our pooled data indicated that the XRCC3 RS861539, TNF-α rs1800629, and IL-6 rs1800795 genetic variants were associated with susceptibility to cervical cancer globally. However, the MTHFR rs1801133, IL-12B rs3212227 and TLR9 rs352140 variants were not associated. Larger and more rigorous studies among different ethnicities are needed to further evaluate these associations with cervical cancer.Seyedeh Fatemeh Parsaeian, Fatemeh Asadian, Mojgan Karimi-Zarchi: conceptualization, investigation. Seyed Alireza Dastgheib, Sepideh Setayesh: Software, original draft preparation. Atiyeh Javaheri, Razieh Sadat Tabatabaie: Investigation. Fatemeh Asadian: Investigation, writing. Fatemeh Asadian, Hossein Neamatzadeh: Methodology, software. Hossein Golestanpour, Hossein Neamatzadeh: Formal analysis, investigation. Hossein Golestanpour: Project administration. Atiyeh Javaheri, Razieh Sadat Tabatabaie: Writing, reviewing, editing.This article does not contain any studies with human participants or animals performed by any of the authors.Not applicable for this manuscript.The datasets generated during and/or analyzed during this study are available from the corresponding author on reasonable request.The authors declare that they have no conflict of interest.


Background: Associations between irregular diet and the risk of esophageal cancer remain unclear. The current meta-analysis was performed to determine whether the presence of irregular diet increases the risk of esophageal cancer.Methods: The data from PubMed, Cochrane Libraries, and Embase up to 23 January 2022 were included in our analysis to identify studies that investigated associations between irregular diet and the risk of esophageal cancer. Summary relative risk (RR) and 95% confidence intervals (CIs) were calculated using a random-effects model.Results: Five cohort studies and one case-control study investigating associations between irregular diet and the risk of esophageal cancer were included. None of the articles demonstrated publication bias. The summary RR was 4.181 . In the subgroup analysis, we found significant heterogeneity in the Non-disease-causing group, nurse group and Asian group. The above three that produce heterogeneity may be the source of heterogeneity in the results of this study.Conclusion: The current meta-analysis indicates that irregular diet increase the risk of esophageal cancer.Trial registration: (https://www.crd.york.ac.uk/prospero/),  Esophageal cancer (EC) is the one of the most commonly diagnosed cancers and the sixth most common cause of cancer-related death worldwide . GeneticIt is controversial that whether irregular diet increase the risk of EC because of a lack of high-quality prospective studies. Thus, A meta-analysis investigating the relationship between irregular diet and EC is warranted is necessary. The vast majority of the existing systematic reviews are on the relationship between specific dietary patterns and esophageal cancer, or the relationship between dietary patterns and other cancers. However, to date no review or meta-analysis has investigated whether there is a relationship between irregular diet and EC risk. Furthermore, cohort studies and case-control studies investigating irregular diet and risk of esophageal cancer are rare. In the current cohort study, as well as in a case-control study, Lene Mellemkjaer et al. showed that anorexia increases the risk of esophageal cancer , Yin ZhaThis meta-analysis was conducted in compliance with the PRISMA statement . The pro765PubMed, Cochrane Libraries, and Embase databases were searched from inception to 23 January 2022. Key search terms included “anorexia nervosa” or “anorexia nervosas” or “eating” or “eating disorder” or “eating disorders” and “esophageal cancer” or “esophageal neoplasm” or “cancer of esophagus” or “cancer of the esophagus” or “esophagus cancer” or “esophagus cancers” or “esophageal neoplasms” or “esophageal cancers” or “cancers” or “cancer”. It is worth mentioning that we used controlled vocabulary (MeSH in Pubmed) and keywords as search terms and the full search strategy was added as supplementary materials . No langThe inclusion criteria included  studies The exclusion criteria included  reviews The Newcastle–Ottawa scale (NOS) was used to evaluate the quality of the studies included . The scaA standardized data extraction sheet was developed in Excel (Microsoft Corporation). Two investigators (Jiayun Guan and Xixin Pan) independently extracted the first author, year, location, study design, sample size, sex, age, rage of BMI, type of Esophageal cancer, causes of irregular diet, sample size, RR, and 95% confidence intervals in the fully adjusted model from the studies. Another investigator (Shenghang Ruan) checked the extracted data. Any discrepancies were discussed with Xiaoxiang Rong and the final decision is made by Xiaoxiang Rong.p value and I2 statistic . And inp = 0.011), with moderate heterogeneity  and Esophageal adenocarcinoma (EAC). ESCC is the mainly pathology type in Asian while EAC is the mainly pathology type in western countries . It is iThe present meta-analysis indicates that a irregular diet increases the risk of EC. The analysis had geographical limitations, and more research is needed to facilitate broader generalization of our conclusion.