More and more researchers have reported that dilatation and curettage (D&C) or Pipelle had low accuracy, high misdiagnosis, and insufficient rate. Endometrial cytology is often compared with histology and seems to be an efficient method for the diagnosis of endometrial disorders, especially endometrial cancer. We report a case of misdiagnosed endometrial cancer by D&C, but with a positive cytopathological finding. Following that, a meta-analysis including 4,179 patients of endometrial diseases with cyto-histopathological results was performed to assess the value of the endometrial cytological method in endometrial cancer diagnosis. The pooled sensitivity and specificity of the cytological method in detecting endometrial atypical hyperplasia or cancer was 0.91[95% confidence interval (CI) 0.74–0.97] and 0.96 (95% CI 0.90–0.99), respectively. The pooled positive likelihood ratio and negative likelihood ratio was 25.4 (95% CI 8.1–80.1) and 0.10 (95% CI 0.00–0.30), respectively. The diagnostic odds ratio which was usually used to evaluate the diagnostic test performance reached 260 (95% CI 36–1905). So we recommend that D&C and Pipelle are still practical procedures to evaluate the endometrium, cytological examinations should be utilized as an additional endometrial assessment method. Endometrial cancer is becoming the primary reason of female deaths of genital track cancer in developed countries . DilatatHere, we report a case of misdiagnosed endometrial cancer by D&C, but with a positive cytopathological finding. The patient has provided her written informed consent for the publication of this manuscript and any identifying images or data. After searching on PubMed, we believe it is the first case report of a misdiagnosis of endometrial cancer detected by cytopathology. Following this, a random-effects meta-analysis including 4,179 patients with both cytopathological and histopathological results was performed to assess the value of the endometrial cytology method in the diagnosis of endometrial atypical hyperplasia or cancer.A 60-year-old post-menopause female, from Baoji City of the Shaanxi province in China, went to a local hospital complaining of abnormal uterine bleeding for 2 months. No high risk factor for endometrial cancer was observed, such as genetic factors, obesity, diabetes, a history of tamoxifen use and so on. Curettage was performed with a histopathological diagnosis of complex hyperplasia endometrium. No medicine or therapeutic curettage was effective for her with a continued bleeding. Her type B ultrasound in Shaanxi Provincial People's hospital showed a 0.8 cm-thick endometrium. Then, she turned to the First Affiliated Hospital of Xi'an Jiaotong University for further treatment. After written informed consent, she volunteered to get cytological endometrial samplings by Li Brush  for cytological examination before D&C. Her histopathological report revealed that papillary epithelial hyperplasia was found, and cancer was a concern according to the structure of tissue but could not be diagnosis due to insufficient tissue . MeanwhiWe searched the PubMed and Embase databases with the heading terms and keywords as “cytology” and “endometrial” from Jan 1, 1995 to June 1, 2018. Then, the results were manually selected for studies to include and repeatedly checked by a second investigator. We searched the full-text articles about the comparison of cytological results and the histological results in endometrial samples.All candidate studies were evaluated and extracted by two independent investigators. Inclusion criteria: (1) patients were diagnosed by histopathological and cytological examination; (2) the histopathological results were paired with cytological results; (3) sufficient information was provided to conduct a statistical analysis; (4) endometrial cells were sampled by endometrial brushes; (5) studies were limited to human trials and published in English. Exclusion criteria include: (1) news, abstracts, case reports, letters, commentaries, and reviews studies; (2) other kinds of endometrial cells sampler like endometrial aspiration cytology; (3) different cytopathology report formats with others, that made it hard to re-group and analyze; and (4) studies with different positive result definition or duplicate data.We set atypical hyperplasia and endometrial carcinoma as the positive results and the others as the negative results, including normal endometrium, non-atypical hyperplasia, endometrial polyp, simple endometrial hyperplasia, complex endometrial hyperplasia and so on.Two investigators separately extracted the following information from each research: the name of the first author, year of publication, cytological sampling method, cytological specimen preparation, histological sampling method, number of patients enrolled, and true positive (TP), false negative (FN), false positive (FP), and true negative (TN) results. Any discrepancies between the two investigators were discussed by all the authors.A quality assessment of eligible studies was evaluated using the quality assessment of studies of diagnostic accuracy included in systematic reviews-2 (QUADAS-2). There were 13 questions : (1) Was a consecutive or random sample of patients enrolled? (2) Was a case–control design avoided? (3) Did the study avoid inappropriate exclusions? (4) Were the cytopathological diagnoses interpreted without knowledge of the results of the gold standard ? (5) whether the blind method was used for pathologists? (6) Was the histopathological diagnosis likely to correctly classify the target condition? (7) Was there an appropriate interval between the cytological sampling and histological sampling? (8) Did all patients receive the histopathological diagnosis? (9) Were all patients included in the analysis? (10) Whether the diagnostic test steps were detailed? (11) Were there concerns that the included patients and setting do not match the review question? (12) Were there concerns that the target condition as defined by the gold standard does not match the question? (13) Were there concerns that the index test, its conduct, or its interpretation differ from the review question?P < 0.05 was considered a significant publication bias. Statistical heterogeneity was detected by a Q test and an inconsistency index (I2), with significant heterogeneity set at P ≤ 0.05 and I2 > 50%. If there was no significance in heterogeneity (P > 0.05), a fixed effects model was chosen. If it was the opposite (P < 0.05), a random effects model was chosen.The publication bias was checked by a Deeks funnel plot, and According to TP, FN, FP and TN results, we calculated sensitivity, specificity, positive predictive value (PPV), negative predictive value (NPV), positive likelihood ratio (PLR) (>10 suggested strong concordance), negative likelihood ratio (NLR) (<0.1 suggested strong concordance), diagnostic odds ratio (DOR). PLR was calculated as: positive likelihood ratio = sensitivity/(1–specificity). NLR was calculated as: negative likelihood ratio = (1–sensitivity)/specificity. DOR was estimated by the Mantel-Haenszel formula. All statistical analyses, including 95% Confidence Interval (CI) were performed using STATA software  with the Midas module.After searching on PubMed and Embase, 9 of 4,182 studies were included in meta-analysis. Our analysis included 9 eligible studies, which were shown in We assessed the quality of eligible studies by QUADAS-2 and found that the quality of all the studies was good .The pooled sensitivity and specificity of the cytological method in detecting endometrial atypical hyperplasia or cancer was 0.91 (95% CI 0.74–0.97) and 0.96 (95% CI 0.90–0.99), respectively . The pooI2 values of pooled sensitivity and specificity were 96.53  and 98.29 , which indicated a statistically significant heterogeneity. A sensitivity analysis was performed to assess the influence of each study, in which each individual study was removed each time. No significant change or reversal of result was found  for sensitivity and 0.93  for specificity. While in subgroup sample size ≥300, the sensitivity was 0.93  and specificity was 0.98 . Subgroup 2: the corresponding values of the studies of European countries showed the sensitivity of 0.96  and specificity of 0.99  and in other countries were 0.84 and 0.92 , respectively .P = 0.60).Deeks funnel plot asymmetry test was conducted to evaluate publication bias in this study , which sGiven the PLR and NLR, the cytological detection method of endometrial atypical hyperplasia or cancer was located in the left upper quadrant , indicatIt is currently estimated, that 60 million cervical cytology examinations are performed every year in the United States . CytopatHistological (D&C and Pipelle) and cytological diagnosis are two classes of endometrial sampling modalities. Both D&C and the most golden standard of evaluating the endometrium, hysteroscopic-guided uterine biopsy, are painful, expensive, and requires dilatation and anesthesia –26. InsuPipelle, as another widely used endometrial biopsy apparatus, is safe, cost-effective, and easily preformed . A meta-Both D&C and Pipelle have their limitations in detecting endometrial cancer. Hysteroscopic guided biopsy showed a high diagnostic accuracy for endometrial cancer diagnosis (estimated sensitivity of 82.6% and specificity of 99.7%), data from a meta-analysis over 9,000 patients , but it Endometrial cytology examination may be an inevitable method for endometrial cancer screening and a combined diagnostic procedure. It might have been hampered by the frequent presence of excess blood, mucus and overlapping cells and varied endometrium cell morphology with different sex hormone levels. However, liquid-based preparation techniques improve the diagnostic accuracy of endometrial cytology , 36, andOur meta-analysis showed that endometrial cytology had a high diagnostic accuracy and could serve as a test to confirm or exclude endometrial atypical hyperplasia or cancer. The pooled sensitivity and specificity of the cytological method in detecting endometrial atypical hyperplasia or cancer was 0.91 (95% CI 0.74–0.97) and 0.96 (95% CI 0.90–0.99), respectively. Its diagnostic odds ratio reached 260 (95% CI 36–1905). The pooled positive likelihood ratio and negative likelihood ratio was 25.4 and 0.10, respectively. Therefore, we can conclude that the test results of endometrial cytology are very accurate in diagnosing endometrial atypical hyperplasia or cancer.Therefore, we recommend that D&C and Pipelle are still practical procedures to evaluate the endometrium, cytological examinations should be utilized as an additional endometrial assessment method, especially for women at high-risk for endometrial cancer.Additionally, endometrial cytology is inexpensive, tolerated well and can be performed without anesthesia in an outpatient clinic. It is now the most common test for an initial evaluation of endometrial cancer in Japan  and has Many researchers reported a high risk of endometrial cancer with positive cervical cytology , 43. AbnAn important strength of this meta-analysis is that we performed a thorough search for articles on the diagnostic accuracy in women with endometrial atypical hyperplasia or cancer using endometrial cytology. This article has several limitations. First, the risk of missing potentially relevant articles is a concern. Otherwise, the relatively small number of studies and variability in methods did not allow for more standard statistical analyses. Higher sensitivity and specificity could be found in subgroup of studies with sample size ≥ 300 and studies in European countries. However, after the subgroup analyses and sensitivity analysis, no factor showed associated with high heterogeneity. Patient age, menopause or not, different kinds of clinical symptoms, varies of cytological samplers and histological sampling methods might contribute to the high heterogeneity, and further study should approve it with enough data. What's more, the studies that are included in the meta-analysis are performed in symptomatic women. More data are needed before endometrial cytology being an effective screening tool for asymptomatic women with high-risks of endometrial cancer.In conclusion, endometrial cytology is an efficient diagnostic method and could be applied in the diagnosis of endometrial disorders. The diagnostic accuracy of endometrial carcinoma will surely be improved by the combination of cyto-histopathological procedures and vaginal ultrasonography. Moreover, cytological examination, as a proper outpatient procedure, should be advised for endometrial screening, especial for those with high-risks of endometrial cancer.QingW and QiW drafted the manuscript. QingW, LZ, and CS collected the case. QiW, LH, and SM performed the meta analysis. HH performed the pathological figure. QS helped to revise the manuscript. QL conceptualized the study.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.


Segmental colectomy was significantly associated with an increased relative risk of MCC . These mutations have a high penetrance with early onset of colorectal cancer at a mean age of 45 years. The mainstay of surgical management is either a segmental or extensive colectomy. Currently there is no unified agreement as to which management strategy is superior due to limited conclusive empirical evidence available. A systematic review and meta- analysis to evaluate the risk of metachronous colorectal cancer (MCC) and mortality in LS following segmental and extensive colectomy. A systematic review of the PubMed database was conducted. Studies were included/ excluded based on pre-specified criteria. To assess the risk of MCC and mortality attributed to segmental or extensive colectomies, relative risks (RR) were calculated and corresponding 95% confidence intervals (CI). Publication bias was investigated using funnel plots. Data about mortality, as well as patient ascertainment  were also extracted. Statistical analysis was conducted using the R program (version 3.2.3). The literature search identified 85 studies. After further analysis ten studies were eligible for inclusion in data synthesis. Pooled data identified 1389 patients followed up for a mean of 100.7 months with a mean age of onset of 45.5 years of age. A total 1119 patients underwent segmental colectomies with an absolute risk of MCC in this group of 22.4% at the end of follow-up. The 270 patients who had extensive colectomies had a MCC absolute risk of 4.7 In 2012, 694,000 deaths worldwide were attributed to colorectal cancer (CRC) alone [MLH1, MSH2, MSH6 or PMS2 genes) which are responsible for correcting base–base mismatches and insertion/ deletion loops arising during DNA replication and recombination. This subsequently results in defective MMR that drives tumourigenesis. Although certain deletion mutations in a non-MMR gene EPCAM can cause Lynch Syndrome, this is because it affects transcription of MSH2.LS is an autosomal dominantly inherited condition which occurs due to a germline mutation in one or more DNA mismatch repair genes (MMR)  Working group who currently recommend that all patients diagnosed with colorectal cancer, irrespective of age, should be universally tested for molecular features suggestive of LS , 10, 11.Synchronous and metachronous colorectal cancers (MCC) occur more commonly in patients with LS compared to the general population , 13. LS The aim of this systematic review and meta-analysis is to conclusively evaluate the risk of MCC and mortality following a segmental and extensive colectomy in LS.An extensive literature search of the PubMed database from inception (last search November 2016) was conducted. Initially the following terms were searched: “Lynch Syndrome” or “HNPCC” AND “metachronous”. This literature search was independently performed by two researchers (SS and ML).The study titles were examined for potential relevance and the abstract was then reviewed. The full text was retrieved to ensure eligibility using the following inclusion and exclusion criteria.≥ 50 cases with previous colorectal surgical resection for colorectal cancer or cancer prophylaxis.LS or HNPCC patients.Median follow-up ≥ 60 months.Full articles published in English within a peer reviewed journal.Sufficient data was available for analyses to be conducted.Insufficient information within the article for inclusion/exclusion to be established.Study included non-colorectal cancer cases.Study included data on patients without a diagnosis of LS or HNPCC.Case only study.Cell or animal based studies.Review articles.The bibliographies of relevant articles were also inspected for further eligible studies. In the case of any uncertainty regarding study inclusion, another investigator (KM) was consulted as a third person arbiter to assess eligibility.For eligible studies, the following data was extracted using a standardised database: title; first author; publication year; number of male and female subjects; mean age of diagnosis; mean follow up time; total number of patients in each study; total number of patients undergoing segmental; subtotal or total colectomy as well as the total number who developed MCC after each surgery and total mortality after each surgery. Extensive surgery was considered to include those patients who underwent panproctocolectomy, subtotal or Total colectomy.If the required data was unavailable within the full text of the article or any accompanying supplementary material, this was sought through e-mail exchange with the stated correspondent(s).Meta-analyses were conducted to quantify the risk of MCC; risk of mortality attributed to colorectal cancer specifically and risk of mortality overall after a segmental versus an extensive colectomy. Quantitative synthesis involved calculating relative risk (RR) for three or more studies using the following formula: For example, whereby segmental colectomy patients with (a) or without MCC (b) were compared to extensive colectomy patients with (c) or without MCC (d).2 test were used to determine the degree of variation between studies not attributable to chance [2 values ranged from 0% indicating homogeneity to 100% indicating heterogeneity. If the I2 value was between 50 and 100%, the DerSimonian and Laird random effects method was used to generate pooled RRs [2 value was between 0 and 50% then the Mantel–Haenszel fixed effects method was used [Corresponding 95% confidence intervals (CI) were deemed statistically significant if they did not cross 1. Cochrane’s Q statistic and the Io chance , 15. I2 oled RRs . If the was used .Subgroup analysis was performed to investigate if there was a significant difference in risk of MCC after segmental versus extended colectomy in patients with LS with a confirmed germline mutation, and separately assessing risk in those patients who met Amsterdam criteria alone but in whom germline status was uncertain.Sensitivity analysis was achieved by removing each study individually and repeating the analysis. This assessed whether the results were influenced by a single study.Publication bias was assessed using funnel plots and Egger’s test. An asymmetrical funnel plot signified the possible presence of publication bias. Publication bias was quantitatively assessed using Egger’s test, however restricted to meta-analyses containing 10 or more studies as per recommendations outlined by Sterne et al. , 19. A pStatistical analyses were performed using the Metafor package in R (Version 3.3.2) .A search of the PubMed database yielded 181 studies. After screening titles and abstracts 165 papers were excluded and 16 studies were reviewed in full. After thorough review of these studies, ten were eligible for inclusion in quantitative synthesis  which will require lifestyle and dietary adaptation and has been reported to negatively impact up to 50% of patient’s sex lives . AlternaA pan-proctocolectomy would be the optimal choice to eliminate a patient’s risk of MCC, especially if the primary tumour has occurred within the rectum. If a patient opts against this and prefers a total colectomy and ileorectal anastomosis or IPAA, then regular endoscopic surveillance of the remaining rectum would be required due to the risk of a rectal cancer. Where a segmental colectomy is the surgical management of choice regular endoscopic surveillance would be sensible due to the five times greater risk of MCC in such patients.Despite there being a significantly increased risk of MCC, the choice of surgical management has no significant effect on mortality. However, it can logically be assumed that a cancer recurrence, need for repeat surgery and anaesthesia would all increase one’s risk of mortality. Despite our findings contradicting this, the limitations of the mortality meta-analysis must be recognised. There were only 3 studies (548 patients) that included mortality data. Before definitive conclusions are made, the meta-analysis should be repeated when more studies with mortality data becomes available. We therefore suggest that further studies investigating the effect of the choice of surgery on mortality should be conducted and authors should make this data available to allow for a conclusive meta-analysis to be conducted.MLH1 and MSH2 mutations, with only two of the studies reporting mutations for MSH6 and PMS2 mutations as well. Thus the high risk of MCC in the segmental colectomy cohort is largely due to patients with MLH1 and MSH2 mutations, and therefore germline specific information must be taken in to account when counselling patients about the risks of surgery [It appears that patients with higher penetrance forms of LS may be at the greatest risk of MCC. Germline specific data reveal that most of the cases in the meta-analysis were patients with  surgery , 25.The choice of surgical management should also take into account the patient’s age and co-morbidities. Parry et al. showed that there was a cumulative increase in the risk of MCC as time following segmental colectomy increases . The risA systematic review and meta-analysis to ascertain the risk of MCC following a segmental colectomy compared to extensive colectomy has been performed by Heneghan et al.  and AnelFirstly, it is not possible to reproduce the odds ratios calculated for 5 out of 6 of the individual studies within the Heneghan study. 3 out of the 6 studies included in the meta-analysis by Heneghan et al. have also been included by Anele et al. and again the odds ratios calculated for these studies do not correlate. This suggests that they’re may have been an error in statistical analysis of the data by Heneghan at al, thus producing an unreliable overall odds ratio for the risk of MCC. Furthermore Anele and colleagues have made an error when extracting data from the study by Stupart et al. . MortaliAnele and colleagues argue that as the Heneghan study includes patients which meet the Amsterdam criteria but do not necessarily have a confirmed germline mutation, this patient demographic could include patients with familial colorectal cancer type X, which has a lower rate of MCC, thus producing bias . Anele eTo investigate if there is a significant difference in risk between patients which meet the Amsterdam criteria (but have not undergone genetic testing) and those with a confirmed germline mutation we performed a subgroup analysis. The relative risk of MCC in these groups was 3.04 (95% CI 1.46–6.34) and 8.56 (95% CI 3.37–21.73) respectively. Despite the risk being lower within those who meet the Amsterdam criteria, there is no significant difference between the two groups.The CAPP2 study demonstrated that regular high dose aspirin could lower the risk of developing colorectal cancer in those with LS . One mayIn addition to chemoprevention, the role of endoscopy in risk modification is not well established. Current international guidance recommends 1–2 yearly colonoscopic surveillance of LS carriers in order to ensure colorectal cancers are identified and managed promptly . SeveralWe conclude that there is a significantly greater risk of MCC following a segmental colectomy compared to extensive colectomy in LS patients. An individualised approach must be utilised taking into account the age, co-morbidities and genotype of the patient as well as the morbidity and impact on the quality of life the choice of surgical management may have. Patients must be counselled on the findings of this meta-analysis prior to any surgical intervention and informed, using the best available germline-specific information. Despite this study demonstrating no significant effect on mortality, better powered studies are required to investigate this further, and an existing international collaboration may be well placed to answer this and other questions . The cur


Vitamin D deficiency is highly prevalent across the globe. Existing studies suggest that a low vitamin D level is associated with more than 130 outcomes. Exploring the causal role of vitamin D in health outcomes could support or question vitamin D supplementation.We carried out a systematic literature review of previous Mendelian-randomization studies on vitamin D. We then implemented a Mendelian Randomization–Phenome Wide Association Study (MR-PheWAS) analysis on data from 339 256 individuals of White British origin from UK Biobank. We first ran a PheWAS analysis to test the associations between a 25(OH)D polygenic risk score and 920 disease outcomes, and then nine phenotypes  that met the pre-defined inclusion criteria for further analysis were examined by multiple MR analytical approaches to explore causality.The PheWAS analysis did not identify any health outcome associated with the 25(OH)D polygenic risk score. Although a selection of nine outcomes were reported in previous Mendelian-randomization studies or umbrella reviews to be associated with vitamin D, our MR analysis, with substantial study power (>80% power to detect an association with an odds ratio >1.2 for per standard deviation increase of log-transformed 25[OH]D), was unable to support an interpretation of causal association.We investigated the putative causal effects of vitamin D on multiple health outcomes in a White population. We did not support a causal effect on any of the disease outcomes tested. However, we cannot exclude small causal effects or effects on outcomes that we did not have enough power to explore due to the small number of cases. Observational studies have identified associations between vitamin D levels and hundreds of disease outcomes. However, evidence of causality for most of these associations is either lacking or not confirmed by randomized clinical trials or Mendelian-randomization studies.With the Mendelian Randomization–Phenome Wide Association Study design, we explored the causal relationships between vitamin D level and 920 outcomes in the UK Biobank cohort. None of these outcomes was causally associated with vitamin D at a moderate or higher effect size.Small causal effects that we did not have enough power to explore in the present study could be studied in the future with larger sample sizes to achieve sufficient statistical power. Vitamin D status is an important public-health issue due to the high prevalence of vitamin D insufficiency and deficiency worldwide.With large cohorts linked to electronic medical records (EMRs), the Phenome Wide Association Study (PheWAS) design has been proposed as a high-throughput approach to comprehensively evaluate associations between genetic variants and a wide range of phenotypes . The PheWAS method has been proven to be useful in the replication of hundreds of known genotype–phenotype associations as well the identification of new associations.In traditional epidemiological analysis, a causal effect of 25(OH)D on disease outcomes cannot be unambiguously demonstrated due to the known limitations of observational research including unmeasured confounding factors and reverse causality. By using single-nucleotide polymorphisms (SNPs) associated with 25(OH)D in an instrumental variable (IV) analysis, also known as Mendelian-randomization (MR) analysis, we can largely overcome these limitations.In the current study, we first conducted a systematic literature review of previous MR studies on 25(OH)D. Then, we analysed 339 256 individuals from the UK Biobank study by implementing a conventional PheWAS and the Bayesian TreeWAS for a large number of health outcomes based on linked EMRs. Finally, for a selection of outcomes based on the results from PheWAS/TreeWAS, power and evidence from previous studies, we performed further MR analysis.IJE online. Any review articles, non-English articles or conference abstracts were excluded. Studies on the impact of other markers/exposures on 25(OH)D levels were also excluded. References from the included studies were checked to identify any additional relevant studies. The literature search and review process were done in parallel by two authors (X.M. and Y.H.). Data extraction was done by X.M. and then confirmed by Y.H.The main steps of the study are presented in UK Biobank is a very large population-based prospective study established to allow detailed investigations of genetic and non-genetic determinants of the diseases in middle- and old-aged adults.IJE online).In this study, we used the UK Biobank genetic data of 488 378 participants. Genetic quality control was done centrally by UK Biobank.et al. for the PheWAS analysis.et al. for the TreeWAS analysis.We implemented all statistical analyses using R 3.3.2. We used the R package developed by Carroll GC), rs12785878 (NADSYN1/DHCR7), rs10741657 (CYP2R1), rs17216707 (CYP24A1), rs10745742 (AMDHD1) and rs8018720 (SEC23A). We created a genetic score by adding the number of effect alleles carried in each of the six SNPs and weighted based on their effect estimates from the SUNLIGHT GWAS.We created a genetic-risk score for 25(OH)D. In the selection of variants, we used the results from the largest genome-wide association study (GWAS) on 25(OH)D in a White population with a total of 79 336 individuals of European ancestry (the SUNLIGHT GWAS).P-value of <5.44 × 10–5 was regarded as statistically significant based on Bonferroni correction (0.05/920 = 5.44 × 10–5).The association between the genetic-risk score and common confounders was first tested. In the PheWAS analysis, we only included disease groups with more than 200 cases, as suggested by a simulation of power estimates for PheWAS analysis.We also applied the TreeWAS Bayesian analysis. The case–control status of participants was defined by their ICD10 codes . Due to the complexity of converting ICD9 codes into ICD10 codes, records of ICD9 were discarded. The same weighted genetic-risk score was employed to test the association between the score and all ICD10 codes presented in UK Biobank data. In our TreeWAS analysis, nodes with posterior probability (PP) of a non-zero effect >0.75 were considered as significant.IJE online). In addition, we constructed the identical genetic-risk score in 2821 control individuals of the Study of Colorectal Cancer in Scotland (SOCCS)  to estimate the variance of 25(OH)D levels explained by this score and the corresponding F-statistics, given that we did not have individual 25(OH)D-level data from the UK Biobank.For those phenotypes for which there was enough statistical power for a MR study (>80%) and (i) were statistically significant in the aforementioned PheWAS/TreeWAS analysis or (ii) were classified as probable or suggestive in our previous umbrella reviewIJE online). The causal effect of vitamin D has been examined across a range of disease outcomes and a causal role of vitamin D is not supported for the majority of them. Disease outcomes that were ever reported to be causally associated with vitamin D levels include type 2 diabetes (T2D), total adiponectin, diastolic blood pressure (DBP), risk of hypertension, multiple sclerosis, Alzheimer’s disease, all-cause mortality, cancer mortality, mortality excluding cancer and cardiovascular events, ovarian cancer, HDL-cholesterol, triglycerides, high-density lipoprotein, delirium and cognitive functions. However, for some of these outcomes, the evidence across different MR studies is not consistent. Detailed results of the systematic review of previous MR studies are present in IJE online, and IJE online.After applying our inclusion and exclusion criteria, 63 MR studies were included in our systematic literature review (IJE online). The association between the score of six SNPs and common confounding factors is presented in P = 1.30 × 10–17), all other confounding factors were not associated with the score. We then tested the associations between the SNPs and the geographic regions  and found that two SNPs (rs12785878 and rs10745742) were unevenly distributed across the UK. Since we have adjusted for the assessment centre and latitude as covariates, we did not expect the association between score and assessment centre to bias our results.We included 339 256 British White unrelated individuals from the UK Biobank cohort, 53.68% of whom were female. All six SNPs satisfied the Hardy-Weinberg equilibrium test  and nephrotic syndrome  (IJE online). The P-value for the association between the score and vitamin D deficiency was 0.00116 (291 cases), which was the third smallest P-value among all tested associations. We additionally performed a sensitivity analysis without adjustment for BMI (IJE online) and the results showed no difference in the significance of PheWAS associations when compared to that with adjustment for BMI (IJE online).In the conventional PheWAS, we tested associations between the score and 920 outcomes (>200 cases). No associations survived the Bonferroni multiple-testing correction. There were only two phenotypes with a In the TreeWAS, the largest PP for all tested phenotypes was 0.26, which was for the calculus of the ureter . Since a PP >0.75 was needed to take findings from the TreeWAS forward, there were no putative associations found from the TreeWAS analysis.N = 2821). The R2 value was 1.61% and the F-statistic was 45.96, indicating that the genetic-risk score is a strong IV for our MR analysis.P-value of the intercept term for all outcomes was >0.05, indicating no evidence of unbalanced pleiotropy (IJE online) among the variants we used.Based on our eligibility criteria, the following outcomes were selected to be further examined using MR: systolic blood pressure (SBP), DBP, risk of hypertension, risk of T2D, risk of ischaemic heart disease (IHD), BMI, risk of depression, risk of non-vertebral fracture and all-cause mortality. Incorporating self-reported data increased the numbers of cases for some of the outcomes . We firsIn order to investigate the safety and rationale of population-wide measurements to raise vitamin D levels (such as fortification of staple food with vitamin D), we conducted a high-throughput PheWAS/TreeWAS study on more than 920 outcomes in a UK Biobank sample of 339 256 British White individuals, followed by MR analyses for selected outcomes. The large sample size ensured the study was well powered to detect moderate to large causal effects (OR >1.2) for outcomes with more than 9997 cases.The PheWAS/TreeWAS analysis used a weighted genetic score of six SNPs as a proxy of the genetically determined 25(OH)D level and examined its association across a wide range of disease outcomes. In this initial scan, we did not observe any significant associations with adjustment for a number of covariates. Additionally, given that there is much debate in the literature suggesting not to adjust for heritable covariates in genome-wide association studies,We then integrated EMR data with self-reported medical conditions for the final MR analysis and explored the causal effect of vitamin D on SBP, DBP, risk of hypertension, T2D, IHD, BMI, risk of depression, non-vertebral fracture and all-cause mortality. This merged dataset increased the number of cases and re-assigned spurious controls, thus increasing the study power. For outcomes tested in MR, we had statistical power of nearly 100% for detecting a true effect of or larger than 1.2, except for all-cause mortality (85% power). The estimated causal-effect sizes for all outcomes were close to null with narrow confidence intervals, which suggested that there is no moderate to large causal effect of 25(OH)D on the nine tested outcomes.,DHCR7 and CYP2R1) as their IV, which explained only 1.0% of the variance for the 25(OH)D level. However, in our study with 85% power for detecting an effect of 1.2 and with a comparable case size (N = 9830 deaths), we did not observe a causal effect . The association between vitamin D and all-cause mortality needs to be studied by a larger MR or meta-analysis of MR studies.Among these nine outcomes, associations between vitamin D and blood pressure, hypertension, T2D, IHD and BMI were suggestive from previous observational studies and RCTs.Our study has several strengths. This is the first study to investigate the causal effect of vitamin D in a large sample of 339 256 individuals across the whole spectrum of disease outcomes. Taking advantage of the PheWAS design, we tested the association between the 25(OH)D genetic-risk score and a wide spectrum of phenotypes. Then, we applied multiple MR methods, including two-stage MR, IVW MR and Egger’s MR, to explore the causal effect and test the robustness of our findings across multiple methods.F-statistics indicates a robust instrument, the variance explained is low. Moreover, considering participants in the UK Biobank are not representative of the UK population and there is evidence of a ‘healthy volunteer’ selection bias,The study also has some limitations. In our analysis, we only included White individuals residing at a high latitude, which may hinder the generalizability of our findings to other populations. The weights we employed in the score creation were from a meta-analysis of GWAS,,Furthermore, the relationship between the 25(OH)D level and the risk of diseases may be nonlinear. As shown by previous studies, vitamin D supplementation only shows treatment effects among individuals with baseline 25(OH)D levels of no more than 30 nmol/L. When all participants were analysed irrespective of their baseline 25(OH)D levels, there was no treatment effect.Although we incorporated EMR data and self-reported medical conditions in our definition of phenotypes, problems with reporting bias could have occurred in outcome definition. Incorporation of more data, including general-practice data, outpatient data, prescription data and even imaging data, would help improve the validity of case definition in PheWAS studies. We set the minimum case number per phenotype based on a simulation analysis of PheWAS power estimates.Our study suggested that there was no evidence of a large to moderate (OR >1.2) causal effect of vitamin D on a number of health outcomes, particularly for SBP, DBP, the risk of hypertension, T2D, IHD, BMI, depression, non-vertebral fracture and all-cause mortality. Further, larger studies, probably involving the joint analysis of data from several large biobanks, may be needed to investigate smaller causal effects that nevertheless could be important for public health due to the high prevalence of low vitamin D levels in many populations.This work was supported by Cancer Research UK (grant number C348/A18927) to M.D., E.T., H.C., S.F.; Cancer Research UK Career Development fellowship (grand number C31250/A22804) to E.T.; and the National Institutes of Health (grant number R01 HL133786) to W.Q.W.; X.M., X.L., Y.H. and H.W. were supported by the China Scholarship Council.Conflict of interest: None declared.dyz182_Supplementary_DataClick here for additional data file.


C677T point mutation in methylenetetrahydrofolate reductase (MTHFR) gene have been found to be associated with ischemic stroke in general population, while the results seem inconsistent. We aim to assess the association between variant MTHFR C677T variant and increased risk of ischemic stroke and focus on the elderly population.We searched PubMed, Embase, Cochrane Library, and Web of Science for eligible studies. Odds ratios (ORs) were calculated with the two-tailed 95% confidence intervals (CIs) by using a random effects model to evaluate any possible association. Among the Chinese and non-Chinese populations, we conducted a subgroup analysis.P = 0.0067 for CT + TT vs. CC; OR = 1.18, 95%CI 1.01–1.38, P = 0.0333 for CT vs. CC; OR = 1.41, 95%CI 1.14–1.75, P = 0.0016 for TT vs. CC; OR = 1.27, 95%CI 1.05–1.54, P = 0.0145 for TT vs. CC + CT; OR = 1.18, 95%CI 1.06–1.31, P = 0.0023 for T-allele vs. C-allele). Further subgroup analyses in the Chinese population indicated that MTHFR C677T variant was associated with a higher risk of ischemic stroke.The electronic database search yielded 1,358 citations as of December 2017; finally, nine case-control studies involving 3,337 subjects fulfilled our eligibility criteria for inclusion in the study. The pooled results showed that MTHFR C677T variant increased the risk of ischemic stroke . Stroke is the second leading cause of mortality worldwide and is the most common cause of long-term disability . It can Ischemic stroke is a clinical condition characterised by reduced blood supply to parts of the human brain. The inadequate blood supply might initiate the ischemic cascade and further lead to dysfunction of the affected parts of brain tissue . IschemiMTHFR, a folate-dependent enzyme, by converting 5, 10-methylenetetrahydrofolate to 5-methyltetrahydrofolate, plays a critical role in regulating plasma homocysteine levels . A C677T‘single nucleotide polymorphism (SNP)’ or ‘SNP’ or ‘mutation’ or ‘genetic polymorphism’ or ‘variation’ or ‘polymorphism’ or ‘variant’) and  and . In addition, the reference lists of original and review articles on the same topic were searched manually.Eligible studies were to be identified by searching electronic databases  by using the following search terms: (All the studies were examined thoroughly to assess their appropriateness for inclusion. The following criteria were used to select studies: (1) published as an original article with cohort or case-control study design; (2) a diagnosis of ischemic stroke reported in the case group and absence of this disease in the control group; (3) MTHFR C677T genetic variant status was classified for each patient; (4) the mean (median) ages of the subjects were > 65 years in the study groups; and (5) published in the English language. Studies that provided no information on patients’ average age were excluded from the meta-analysis.For the meta-analysis of MTHFR C677T genetic polymorphism and ischemic stroke, the following information was extracted from each study: (1) first author’s name; (2) publication year; (3) study groups and their respective numbers of cases and controls; (4) numbers or percentages of subjects with the CC, CT, and TT genotypes in the cases and controls; and (5) baseline characteristics of the study patients .We used standardised data recording sheets to extract data from the included studies. Two investigators (Z.K. and J.W.) separately performed the screening of study titles and abstracts, full-text reading, and final data extraction. In case of disagreement, the third investigator (JY.W.) was consulted to reach a final agreement. The potential risk for bias was evaluated in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-analysis recommendations . Study qhttp://www.R-project.org/) and the Meta package [I2 value of > 50% [P value of < 0.10 in the Cochrane Q test [The meta-analysis was performed using R 3.4.3  and non-Chinese populations . The subgroup results in the Chinese population similarly showed that MTHFR C677T variant and T-allele expression were significantly associated with elevated risk of ischemic stroke as follows: CT + TT versus CC, OR 1.32, 95%CI 1.09–1.59, Fig. A further pre-specified subgroup analysis was conducted in the Chinese (P = 0.402). We performed a meta-regression analysis including ethnic group as a factor and didn’t found any significant difference among five panels compared (data not shown).The funnel plots did not show any evidence of obvious asymmetry Fig. . We furtTo the best of our knowledge, this study is the first meta-analysis to focus on a specific elderly population to assess the risk of ischemic stroke and MTHFR C677T variant. The large-sample pooled analysis of nine homogeneous case-control studies demonstrated an association between MTHFR C677T variant and increased risk of ischemic stroke as follows: a 23% (95% CI 6–43%) higher risk of ischemic stroke was observed in the MTHFR C677T mutation group than in the control group with no mutation, and up to 32% (95% CI 9–59%) higher risk was observed specifically in the Chinese subgroup.Hyperhomocysteinaemia is emerging as a potential risk factor of cardiovascular diseases, including cerebral stroke, and is considered modifiable. Both genetics and environmental factors  affect homocysteine level . As one In China, with a population of 1.4 billion, the annual stroke mortality is approximately 1.6 million, and approximately 43 to 79% of all stroke cases are ischemic , 42. ThiThis meta-analysis showed that MTHFR C677T variant may be associated with increased risk of ischemic stroke in the elderly, specifically a significant difference among the Chinese population and a similar trend among the non-Chinese population as the non-Chinese subset is smaller than the Chinese subset which can be explained by a difference in power. However, we agree that this meta-analysis has several potential limitations. First, the study was based on secondary study-level data. All the included studies in our meta-analysis met the required mean (median) age of ≥65 years for study subjects, but this is not to say that each subject from the included studies was aged > 65 years. Second, the homocysteine concentration was measured only in two of the included studies , 30, whiIn conclusion, our findings showed that MTHFR C677T variant may contribute to increased risk of ischemic stroke in the elderly population; T-allele increases risk for stroke in the pooled sample. This association was statistically significant in the Chinese cohorts and showed a similar trend in the non-Chinese cohorts. For elderly people, MTHFR C677T might be a promising biomarker for the early detection and prediction of the prognosis of ischemic stroke. For more solid evidence, however, well-controlled prospective studies are warranted in the future.


Dairy products contain both beneficial and harmful nutrients in relation to cardiometabolic diseases. Here, we provide the latest scientific evidence regarding the relationship between dairy products and cardiometabolic diseases by reviewing the literature and updating meta-analyses of observational studies.We updated our previous meta-analyses of cohort studies on type 2 diabetes, coronary heart disease (CHD), and stroke with nine studies and confirmed previous results. Total dairy and low-fat dairy (per 200 g/d) were inversely associated with a 3–4% lower risk of diabetes. Yogurt was non-linearly inversely associated with diabetes . Total dairy and milk were not associated with CHD (RR~1.0). An increment of 200 g of daily milk intake was associated with an 8% lower risk of stroke.The latest scientific evidence confirmed neutral or beneficial associations between dairy products and risk of cardiometabolic diseases.The online version of this article (10.1007/s13668-018-0253-y) contains supplementary material, which is available to authorized users. Diet-related cardiometabolic diseases such as type 2 diabetes mellitus, coronary heart disease (CHD), and stroke result in a large global health burden, accounting for over 17 million deaths in 2010 . It is iDairy products are worldwide increasingly consumed as indicated by recent tabulations by the International Dairy Federation. From 2006 to 2013, global steep increases in per capita dairy product consumption were found. Especially, Asia, Africa, and Latin America are growing markets for dairy consumption . Of note, full-fat milk, still contains considerably less fat than low-fat cheese per 100 g product. Cheese contains 50% less water than milk and is therefore more nutrient dense than milk or yogurt. Cheese is relatively high in saturated fat, with low-fat cheese containing 8 g more fat than low-fat milk or yogurt per 100 g product. Dairy contains not only various nutrients beneficial for cardiometabolic health such as calcium, potassium, phosphorus, different vitamins such as B2, B12, D and K2, but also harmful nutrients such as sodium, saturated fat, and added sugars [Dairy products such as milk, cheese, and yogurt are nutrient-dense as described in Table d sugars –15.TableBesides high nutrient and energy densities, dairy is heterogeneous comprising many products. For example, butter is a type of dairy, yet it is often grouped with fats and oils because of the role butter has in the diet. Initially in scientific research milk and total dairy, including a variety of products, were grouped together and as such compared between studies –18. In dMany reviews and meta-analyses summarizing associations between dairy products and cardiometabolic diseases were published in the past 20 years ••, 25••.This review provides the latest scientific evidence on dairy products in relation to cardiometabolic diseases . In addition to reviewing a large range of meta-analyses of prospective cohort studies, we identified recent cohort studies to update our previous meta-analyses on dairy in relation to type 2 diabetes, CHD, and stroke ••, 25••.Search syntaxes were similar to our previous meta-analyses ••, 25••.We extracted descriptive study data as well as ranges of intake, medians or midpoints, numbers of subjects and stroke events, person-years at risk, and relative risks (RRs) with the corresponding 95% CIs for each category of dairy intake. Portion sizes of dairy products to convert these into grams per day, were generally reported in the included studies. If dairy intake was only reported in servings, without the actual portion size, we used portion sizes of 177 g for total, low-fat, and full-fat dairy; 244 g for total, low-fat and full-fat milk; 244 g for yogurt; and 43 g for cheese to estimate grams per day. These were mainly UK-  and US- 2 statistic [We performed meta-analyses for type 2 diabetes on total dairy, low-fat dairy, and yogurt, for CHD on total dairy and milk, and for stroke on total dairy, milk, full- and low-fat dairy, for which the most interesting findings were reported in our previously published meta-analyses ••, 25••.tatistic . StatistAn overview of all recent reviews and meta-analyses, and prospective cohort studies is summarized in Supplementary Table The evidence on the association between dairy products and type 2 diabetes from meta-analyses was summarized recently by Drouin-Chartier et al. •• and YuThe meta-analysis by Aune et al.  included2 = 74%, n = 21 studies). In subgroup analyses, the inverse association with total dairy and diabetes was observed only in Asian and Australian studies , but not for American and European studies . This was also observed by Gijsbers et al. [As indicated above, dairy has been the focus in a large number of reviews and meta-analyses. Each one included a different perspective on the topic, with different numbers of studies included. Gijsbers et al. •• includs et al. ••. Moreos et al.  for studs et al. . In anots et al. , a smalls et al. , 54, 55 s et al. , 57.I2 = 66% for total dairy, 68% for low-fat dairy, 73% for yogurt, all p values < 0.05). In subgroup analyses, we found a stronger inverse association for total dairy and total milk consumption in Asian populations , compared to a null association in European populations. This may have been due to confounding adjustments (less extensive in the Asian cohorts than in the European cohorts). We updated our meta-analyses for type 2 diabetes with four cohort studies [I2 = 60%)   was the most complete meta-analysis so far, and included 22 prospective cohort studies comprising 579,832 individuals and 43,118 type 2 diabetes cases ••. In su studies –30, 58 f studies •• and siIn recent years, several meta-analyses or reviews of meta-analyses have published results on dairy intake or specific types of dairy and CHD •, 61, 62We updated our previous meta-analyses for total dairy and milk in relation to CHD •• with fIn recent years, several meta-analyses or reviews of meta-analyses have published results on dairy intake or specific types of dairy and incident stroke , 63••. IP = 0.004; I2 = 86%). RRs were 0.82 (95% CI 0.75–0.90) in East Asian and 0.98 (95% CI 0.95–1.01) in Western countries  with less but still considerable heterogeneity within the continents. Cheese intake was marginally inversely associated with stroke risk . Risk reductions were maximal around 125 g/day for milk and from 25 g/day onwards for cheese. Based on a limited number of studies, full-fat milk (n = 4 studies) was directly associated with increased stroke risk, whereas full-fat total dairy (n = 6), as well as low-fat dairy (n = 7) was inversely associated.In 2016, we pooled 18 prospective cohort studies from 11 countries with 8 to 26 years of follow-up that included 762,414 individuals and almost 30,000 stroke events based on a search up to October 2015 ••. An inI2 = 64%). The largest risk reduction was observed at approximately 40 g/d. This non-linear association was also observed in our meta-analysis [The meta-analysis by Chen et al. , focusedanalysis ••, althoI2: 50%; n = 11).Drouin-Chartier et al. (2016) summarized meta-analyses (not including ours) on various dairy products and stroke ••. ThreeIn line with our findings ••, DrouiYu and Hu •• presenI2 = 85%). RRs were 0.82 (95% CI 0.75–0.89) in East Asian and 0.98 (95% CI 0.95–1.01) in Western countries   Fig. . Risk reScientific evidence from prospective cohort studies to date has suggested null or (weak) inverse associations between dairy intake and risk of type 2 diabetes, CHD, and stroke. This review, adding nine new cohort studies to existing meta-analyses, further confirmed these associations. These results should be placed in the context of observed and unexplained heterogeneity. We presented robust findings in line with previous meta-analyses ••, 25•• Some differences between meta-analyses , 45 wereJust before finalizing this review, the PURE cohort investigators published dairy data in relation to 9-year incident all-cause mortality, major CVD events, and  stroke in 21 countries from five continents  . This inWe and others described potential mechanisms previously ••, 24••.The current discussions and confusion on health effects of fatty acids have contributed to a shift of thinking from a nutrient-based approach towards a food-based approach. Dietary guidelines traditionally mainly focused on saturated fat in dairy products and its low-density lipoprotein raising effects . HoweverButter is the most SFA dense food (50% SFA). It does not contain milk fat globule membranes (MFGM), as opposed to other types of dairy. A recent RCT showed that, in contrast to milk fat without MFGM, milk fat enclosed by MFGM does not impair the lipoprotein profile . A systeAll evidence of dairy in relation to disease endpoints has been derived from observational studies, and residual confounding remains an issue in this type of research. There is evidence that milk and yogurt intake is related to healthy behaviors  and furtTo study causality, additional evidence from RCTs and other types of designs such as Mendelian randomization studies are needed. Drouin-Chartier et al. carried out a comprehensive narrative review on the effects of dairy foods, irrespective of fat content, on cardiometabolic risk factors ••. That Mendelian randomization analyses have been more widely used to assess potential causal estimates of environmental risk factors with health outcomes. This type of analysis has the advantage over observational studies of minimizing confounding by using genetic markers as instrumental variables of environmental risk factors and therefore assessing causality. A recent Mendelian randomization analysis • using 2This systematic review showed neutral or beneficial associations between specific dairy products and risk of type 2 diabetes, CHD, and stroke. In updated meta-analyses, higher milk intake was inversely associated with risk of stroke, but not with risk of CHD or type 2 diabetes. Stronger inverse associations for milk and stroke were found in Asian vs. Western populations, but this has to be investigated further with more Asian population samples. Cheese was not related to diabetes, conflicting with some of the earlier meta-analyses, depending on which studies were included. Most striking results were found for yogurt and type 2 diabetes, but this finding could also be due to more healthy behaviors which are associated with yogurt consumers . Future Supplementary Fig. 1Forest Plot Low-fat dairy intake and risk of type 2 diabetes (JPG 102 kb)Supplementary Fig. 2Forest Plot milk intake and risk of coronary heart disease (JPG 89 kb)Supplementary Fig. 3Ding’s Spaghetti plot for milk intake and risk of stroke. Light blue lines represent Western countries and brown lines represent Asian countries. (JPG 54 kb)Supplementary Table 1(DOCX 56 kb)


Therefore, this study may serve a better position as a citable article.The systematic review and meta-analysis study “Increased risk of second cancers at sites associated with HPV after a prior HPV-associated malignancy, a systematic review and meta-analysis”, presented by Gilbert et al. and published in BJC, has been read by us with interest.3 This issue needs to be addressed as a major limitation of this study.A question that presents itself upon viewing the data presented in the study is the choice of Gilbert et al. to conduct a meta-analysis when a number of included studies have very small sample sizes. The problems that arise upon choosing to include studies with small sample sizes is self-evident in Figure 3 of Gilbert et al.’s paper, where we observed that the number of studies .5The abovementioned issues are exacerbated when we observe that the authors have used the term SIRs interchangeably with HRs, particularly in the forest plots  where they continue to describe the SIRs in the legends of each figure, while the figures themselves seem to represent HRs. The estimated or pooled effect size of the risk of HPV-associated cancer after prior diagnosis is SIR in this meta-analysis. In statistical terms, HRs are markedly different from SIRs, as the former denotes hazard rates corresponding to the conditions described by two levels of an explanatory variable, while the latter describes the ratio of the observed number of cases to the expected number of cases, and therefore, must not be used interchangeably.8It is also significant that the study did not evaluate the presence of publication bias in its included studies. Despite its absence being a legitimate threat to the validity of any meta-analysis study, few meta-analysis studies include it in their analysis, even though methods to assess it has been addressed in previous studies, elaborating on its importance.9Another point of consideration would be the test of between-study heterogeneity. The authors have used the Higgins I-squared statistic to analyse the meta-analysis data. However, the I-squared statistic may not be sufficiently informative as it does not consider the threshold effect. We want to suggest that Kendall’s Tau-squared statistic to be included. This Tau-squared statistic serves as the estimated variation of heterogeneity between the effects for test-accuracy observed between studies, while also considering the threshold effect.These issues are not only relevant to Gilbert et al.’s study, but also for all systematic reviews and meta-analysis which may seek to follow up on similar lines of study.


To drive a more precise estimation, we carried out this meta-analysis. The literature was searched from MEDLINE (mainly PubMed), Embase, the Cochrane Library, and Google Scholar databases to identify eligible studies. A total of 89 studies were included. The results showed that miR-146a C/G was significantly associated with increased HNC risk in dominant model  =1.088, 95% confidence interval (CI) =1.002–1.182, P=0.044). However, no cancer risk was detected under all genetic models. By further stratified analysis, we found that rs4919510 mutation contributed to the risk of HNC amongst Asians under homozygote model , and dominant model . Simultaneously, in the stratified analysis by source of controls, a significantly increased cancer risk amongst population-based studies was found under homozygote model, dominant model, recessive model, and allele comparison model. However, no significant association was found in the stratified analysis by ethnicity and source of control. The results indicated that miR-146a C/G polymorphism may contribute to the increased HNC susceptibility and could be a promising target to forecast cancer risk for clinical practice. However, no significant association was found in subgroup analysis by ethnicity and source of control. To further confirm these results, well-designed large-scale case–control studies are needed in the future.Several studies have evaluated the association of  Cancer, although an age old disease, still poses a formidable challenge to researchers and clinicians. Little is known about its initiation, sustenance, progression and metastasis, and resistance and remission. Due to its morbidity and mortality, cancer is one of the most dreaded diseases and the related fatalities are majorly attributed to delayed diagnosis and treatment. Head and neck cancer (HNC), the sixth most frequent kind of cancer worldwide, is a group of biologically similar cancers that originate from head and neck regions such as oral cavity, pharyngeal cavity, and larynx . MultifamiRNAs represent a class of evolutionarily conserved, endogenous, single-stranded, non-coding RNA molecules of ~20 nts that regulate gene expression by degrading mRNAs or suppressing translation. miRNAs have been implicated in a wide range of physiologic and pathologic processes, including development, cell differentiation, proliferation, apoptosis, and carcinogenesis ,5. AccummiR-146a C/G is located in the stem region opposite to the mature miR-146a sequence, which is suspected to have an effect on tumor immune responses and ultimately the development of cancer. In recent years, the polymorphism rs2910164 in miR-146a has attracted wide attention and many studies have been published to explore the association between SNPs of miRNAs and susceptibility to various cancers. But the results were not conclusive and consistent. Since SNPs in miRNAs are closely associated with head and neck cancer (HNC) susceptibility, it is necessary to assess whether these SNP polymorphisms are the risk factors for HNC. It is reported that meta-analysis is a well-established method for combining all the results from the available published information to produce a single estimate for quantitating gene–disease associations more precisely to increase the statistical power [miR-146a C/G polymorphism for HNC susceptibility.Single nucleotide polymorphisms (SNPs) are a type of common genetic variations associated with population diversity, disease susceptibility, drug metabolism, and genome evolution . SNPs maal power . Thus, wmiR-146a’ or ‘rs2910164’ and ‘head and neck cancer’ or ‘cancer’ or ‘tumor’ or ‘carcinoma’ and ‘polymorphism’ or ‘SNPs’ or ‘variation’. All eligible studies published in English were retrieved, and their bibliographies were checked for additional relevant publications. Review articles and bibliographies of other identified relevant studies were searched by hand to identify any additional eligible studies.A comprehensive electronic search was performed to identify articles published up until 12 November 2016 in MEDLINE (mainly PubMed), Embase, the Cochrane Library, and Google Scholar using the following search terms: ‘miR-146a C/G polymorphism and susceptibility to HNC and overall cancer; (ii) sufficient published data for calculating odds ratios (ORs) with corresponding 95% confidence intervals (CIs); (iii) full-text manuscript; and (iv) only the most recent or complete study reporting on the same population of patients was included. Exclusion criteria included: (i) reviews, other meta-analyses, comments, letters, and editorial articles; (ii) not a case–control study; and (iii) no usable data reported.Studies included in this meta-analysis had to meet all of the following criteria: (i) case–control study evaluating the association between Information regarding the following aspects was independently retrieved from each study by two reviewers: the first author’s surname, year of publication, country of origin, ethnicity, study design, total number of cases and controls, source of cases and controls, detected sample, genotyping methods, allele and genotype frequencies of cases and controls, and evidence of Hardy–Weinberg equilibrium (HWE) in the controls. In studies including subjects of more than one ethnicity, genotype data were extracted separately for each ethnic group. Data from one publication may contain more than one seperate case-control studies. Any discrepancies between the reviewers were resolved through discussion to reach a consensus.miR-146a C/G polymorphism and the risk of HNC and overall cancer. Five genetic variation models were analyzed: homozygote model (CC compared with GG), heterogeneity model (GC compared with CC), dominant model (CC + GC compared with GG), recessive model (CC compared with GC + GG), and allele comparison model (C compared with G). P-value of HWE in control group of each study was calculated by χ2 test and P<0.05 presented a state of disequilibrium [χ2-based Q-statistic [P-value of heterogeneity was less than 0.05) and fixed-effects model  were used [P-values <0.05 were considered statistically significant.We used crude ORs with 95% CIs to explore the association between ilibrium . We alsotatistic . Both raere used ,21. Sensere used ,23. All miR-146a C/G genotype in all studies were in accordance with HWE in the control group. No significant differences were found between cases and controls with respect to gender and age distributions. The modified quality scores of all studies ranged from 9 to 16, with 71% (5/7) of the included studies classified as high quality (≥12).The characteristics of all included studies are summarized in A total of 721 articles were retrieved after the first search in PubMed, Embase, the Cochrane Library, and Google Scholar. Selection following the specified criteria eliminated 632 studies, leaving 89 individual studies –103. ThemiR-146a C/G polymorphism and the risk of HNC under homozygote, heterozygote, recessive, and allele comparison model. There is no significant association between miR-146a C/G polymorphism and the risk of HNC under homozygote model , heterozygote model , dominant model , recessive model , and allele comparison model , so that random-effects model was used in all genetic models. Overall, significant association was not identified in all genetic models  did not identify obvious asymmetry in any of the comparison models, and plot symmetries are evidenced by analysis  and 3.miR-146a rs2910164 polymorphism was not excessively influenced, suggesting the stability and reliability of the results in the present meta-analysis (not shown).We performed sensitivity analysis by sequential omission of individual studies, and the results showed that the significance of the pooled ORs for miR-146a C/G gene have drawn increasing attention in cancer etiologies, and altered expression levels have been observed in inflammatory diseases as well as in cancers [miR-146a C/G polymorphism is associated with HNC risk. This risk is significant amongst the individuals with a dominant genotype model. In the stratified analysis by ethnicity, significant analysis was detected amongst Asians under homozygote and dominant model, while no association was found amongst Caucasians under all genetic models. Furthermore, significant association was found in population-based studies under homozygote, dominant, recessive, and allele comparison models. However, no significant association was detected in hospital-based studies under all genetic models. Moreover, no significant association was found between this gene polymorphism and overall cancer risk. Furthermore, in the stratified analyses by ethnicity and source of control, no significant association was detected in the subgroup analyses of source of control.It is well known that genetic mutations are responsible for cancer occurrence . SNPs, a cancers ,108. ThemiR-146a C/G polymorphism and HNC risk, and the most comprehensive one to date to explore the association between this gene polymorphism and overall cancer risk. Nevertheless, our meta-analysis also has some limitations common to these types of studies. First, relatively large heterogeneity was observed across all the studies involved despite the use of strict criteria for study inclusion and precise data extraction. So, we performed subgroup analyses to explore the possible source of heterogeneity. Second, the majority of subjects included in this meta-analysis were mainly Caucasians and Asians. Thus, the inherent genetic and geographic differences require more data from different ethnic group to increase the statistical power. Third, the low sample size in some of the included studies likely influences the statistical power for evaluating the association between the miR-146a C/G polymorphism and HNC risk, especially in subgroup analyses. Fourth, lack of original data from the reviewed studies limited our further evaluation of potential interactions, considering that gene-to-gene and gene-to-environment interactions might modulate cancer risk. As a result, a more precise analysis stratified by variable host factors could be performed. Last, although the results for publication bias were not statistically significant, publication bias may still exist, because only published studies were included in this meta-analysis.To the best of our knowledge, the present study is the first and most comprehensive one to date to assess the relationship between miR-146a C/G polymorphism more is likely contribute to the susceptibility to HNC, and overall cancer risk. Further well-designed studies with large sample size are needed to confirm these findings.In conclusion, the meta-analysis presented here indicates that


The associations between childhood acute lymphoblastic leukemia (ALL) and several factors related to early stimulation of the immune system, that is, farm residence and regular contacts with farm animals  or pets in early childhood, were investigated using data from 13 case–control studies participating in the Childhood Leukemia International Consortium. The sample included 7847 ALL cases and 11,667 controls aged 1–14 years. In all studies, the data were obtained from case and control parents using standardized questionnaires. Pooled odds ratios (ORs) and 95% confidence intervals (CIs) were estimated by unconditional logistic regression adjusted for age, sex, study, maternal education, and maternal age. Contact with livestock in the first year of life was inversely associated with ALL . Inverse associations were also observed for contact with dogs  and cats  in the first year of life. There was no evidence of a significant association with farm residence in the first year of life. The findings of these large pooled and meta‐analyses add additional evidence to the hypothesis that regular contact with animals in early childhood is inversely associated with childhood ALL occurrence which is consistent with Greaves’ delayed infection hypothesis. Acute leukemia (AL) is the most common cancer in children under 15 years of age According to Greaves’ delayed infection hypothesis An additional early source of immune stimulation may be regular contacts with animals in early childhood, as has been hypothesized for allergies Farm residence involves exposure to animals, especially farm animals, and to various organic dusts, toxins, and pesticides. The limited literature (three case–control studies) In this study, we used a unique large set of case–control studies from CLIC to examine whether living on a farm or regular contact with livestock, poultry, and pets in early childhood (which we defined as the first year of life) was inversely associated with childhood ALL.The data for this analyses were provided by principal investigators of 13 CLIC case–control studies conducted in nine countries between 1980 and 2013 ; whether the child had ever visited a farm and the timing and frequency of these visits; and whether the child had regular contact with animals or whether the family had owned any animals with details on time period and type of animal. Stratification variables  and parental characteristics  were also requested, as well as any other variables used for matching and other indicators of socioeconomic status . For ALL cases, histological data were also provided for each study, except BRA_SAOP and CR_CRCLS, for which they were unavailable.All of the studies were approved by institutional ethics committees.Age was defined as age at the reference date (diagnosis for cases and recruitment or questionnaire return for controls) and was categorized in eight classes . Parental education levels were harmonized across all studies and categorized into three levels: none or primary education, secondary education, and college/university degree. Depending on the available data, a proxy for SES was derived from family annual income , parental professional category , parental education level (IT_SETIL), or housing characteristics (CR_CRCLS) and classed in three categories: low, medium, and high.Data regarding farm residence in the first year of life were available in six studies Table . ChildreContacts with livestock were defined as contacts with at least one type of livestock,  in the first year of life, in their residence , or by regular visits . Seven studies  included contacts with cattle and pigs and contacts with sheep were available in all of them except CR_CRCLS Table . Binary Data on contact with poultry in the first year of life were available in six studies. A binary exposure variable was generated, by classing as exposed to poultry children whose parents had reported that they had poultry at their residence in the child's year of birth  or those who were reported to have had regular contact with poultry in their first year of life .exposed to pets children whose parents had reported that they had a pet at their residence in the child's year of birth  or those who were reported to have had regular contact with pets in their first year of life  or in their two‐first years of life (US_NCCLS). For the later study (US_NCCLS), as only data on regular contact with pets in the two‐first years of life were available, we hypothesized that this period would be a good surrogate for exposures occurring in the first year of life. A four‐class variable  was also created to take into account the different combinations of pet contact.For these analyses, a ‘pet’ was defined as either a cat or dog and 12 studies  had these data  were analyzed using SAS v9.3  and R v3.3.1 software. We restricted the analyses to children aged at least 1 year of age, in order to ensure that the ALL cases and controls had the opportunity to have lived or visited a farm or to be exposed to animals during their first year of life. The analyses were performed for overall ALL and by ALL subtypes, B lineage and T lineage ALL. Analyses were also stratified by age  and sex.QI2 statistics I2 statistic was greater than 0, the random‐effect model Study‐specific ORs were estimated for the main exposures of interest and summarized using IPD meta‐analyses with a two‐stage procedure. First, the study‐specific OR and 95% CIs were estimated, using either unconditional or conditional logistic regression, depending on the original study design, and including study‐specific matching variables in the models. The socio‐demographic characteristics significantly associated with both the case–control status and exposure were also included in the study‐specific models. We investigated between‐study heterogeneity by calculating Cochran's Pooled ORs and their 95% CIs were estimated from the pooled individual data using unconditional logistic regression adjusted for age, sex, and a categorical variable denoting study of origin. Maternal education and maternal age at the child's birth were also included in the models, because they were significantly associated with both case–control status and exposure. We assessed two‐way interaction terms between exposure variables  and conducted stratified analyses to determine whether the effect estimates of a single exposure differed by strata of another exposure.Analyses were repeated using a one‐stage meta‐analysis procedure. For each exposure of interest, ORs and their 95% CIs were estimated from the pooled individual data using unconditional logistic regression adjusted for age, maternal education, and maternal age at the child's birth. The models also included a random effect for the intercept and a random slope for the exposure of interest.Robustness of the results was assessed by excluding one study at a time from the pooled analyses. To detect potential selection bias related to study design, we also repeated the analyses with hospital‐based case–control studies excluded  from the analyses. Analyses were also repeated adjusting for SES instead of maternal education and adjusting for daycare attendance and breastfeeding during the first year of life (yes/no), both factors associated with ALL in previous pooled analyses For each exposure of interest, the potential for participation bias was also assessed by estimating the difference in participation between exposed and unexposed controls that would have generated an OR of the observed magnitude, under the assumption of no true association.P‐values were calculated, even though the question was one‐sided.The 95% CI and two‐sided Overall, data from 13 studies with 7847 ALL cases  and 11,667 controls were included in the analyses Table . The casCase mothers were more often less than 25 years of age at the index child's birth  and less educated than control mothers . Case parents also tended to be in the lowest SES category compared to control parents , but there was high between‐study heterogeneity for contacts with sheep  and 0.87 , respectively .The deterministic sensitivity analyses for home pesticide use Table  and pateWe evaluated the possibility that the inverse association with livestock was due to higher participation of exposed controls. For such a bias to generate an odds ratio of the observed magnitude (OR = 0.65), participation fractions would have had to be of 100% for exposed controls and 76% for nonexposed controls .The present study, based on the largest sample to date from 13 case–control studies, suggests that ALL risk is inversely associated with regular contact with livestock and pets in the first year of life. More specifically, we found inverse associations for regular contact with cattle, pigs, sheep, dogs, and cats. In contrast, our findings do not support an association between living on a farm before 1 year of age and ALL.The existing literature about living on a farm and contact with farm animals  or pets is sparse Regarding contacts with pets and farm animals, the publications from the two French studies  were hospital‐based whereas the others were population‐based. In the sensitivity analyses, the results were similar when the two hospital‐based studies were excluded or when each of the other studies was excluded in turn, which suggests that biases inherent to hospital‐based studies or to one study in particular are not likely to explain the associations. Participation in controls was also high in the included studies. Moreover, the sensitivity analysis showed that a participation bias is unlikely to explain the results, as it would correspond to a differential participation of 100% of exposed controls and 76% on unexposed controls.Recall bias regarding the exposure under study might have occurred, but it is not expected to be differential. Unlike parental tobacco or alcohol beverage consumption, contact with animals and residence in a farm are not exposures for which report may be accompanied by guilt and thus lead to systematic under‐reporting in case parents compared to control parents. Nondifferential errors may also have occurred, which could have biased estimates toward the null, particularly for farm residence, although the exposures of interest are not particularly difficult to remember. To reduce nondifferential errors, because of differences in questionnaires between studies, we used the most specific definition for exposure and kept separate the exposures which could not be aggregated. To deal with potential misclassification related to parent's recall, as exposure may have occurred up to 15 years before the interview, we performed stratified analyses on the child's age at diagnosis/interview, which generated similar results. Lack of specificity for the definition we have used for farm residence might also have generated nondifferential errors. Indeed, in the studies included in this analysis, no details regarding the type of farms were available, which could have added to the level of heterogeneity.In regard to the pooled analysis, potential confounding by factors associated with child age and sex was taken into account when adjusting for these variables. Study of origin was also systematically adjusted for, in order to take into account unmeasured potential confounding factors related to differences between studies such as their designs or recruitment periods. Moreover, our models were systematically adjusted for maternal age at child's birth and maternal education, factors which were associated with both case–control status and the exposures under study. When we also adjusted for socioeconomic status, which is more prone to differences in definition between studies than maternal education, the results were similar. Proxies of early immune stimulation associated with ALL occurrence in a previous CLIC analysis were also taken into account as additional adjustment variables and led to similar results. Children in contact with pets may be exposed to insecticides applied to control fleas and ticks, which are suggested to be positively associated to ALL and not likely to explain the present results. Even if the data were not available for each study, potential confounding by pesticide exposure and paternal smoking was investigated and results from deterministic sensitivity analysis suggested that omitting those variables from the main analysis only led to weak bias in the estimates of the exposures under study.The main disadvantage of the pooled analysis is that study‐specific adjustment variables and specific design such as case–control paired matching cannot be taken into account. However, the results were similar for the meta‐analyses in which adjustment was made for study‐specific stratification variables or confounding factors and conditional logistic models were performed to take into account case–control paired matching when necessary.number of animals in addition to the ever/never variables we have used would have helped to interpret our findings. Nevertheless, such data were only available for the FR_ADELE, FR_ELECTRE, and NZ_NZCCS studies, which precluded a robust analysis.Finally, investigating the exposure related to animals in terms of In conclusion, the findings of these large pooled and meta‐analyses bring additional evidence to the hypothesis that regular contact with animals in early childhood is inversely associated with childhood ALL occurrence. This is consistent with Greaves’ delayed infection hypothesis. Nevertheless, biological mechanisms involved in this association are still to be discovered.The authors declare that they have no conflict of interest.Figure S1. Association between acute lymphoblastic leukemia and contact with cattle in the first year of life (yes vs. no), Restricted to children aged ≥1 year, meta‐analysis of 7 studies (1990–2013), Childhood Leukemia International Consortium.Click here for additional data file.Figure S2. Association between acute lymphoblastic leukemia and contact with pigs in the first year of life (yes vs. no), restricted to children aged ≥1 year, meta‐analysis of 7 studies (1990–2013), Childhood Leukemia International Consortium.Click here for additional data file.Figure S3. Association between acute lymphoblastic leukemia and contact with sheep in the first year of life (yes vs. no), restricted to children aged ≥1 year, meta‐analysis of 6 studies (1990–2013), Childhood Leukemia International Consortium.Click here for additional data file.Figure S4. Association between acute lymphoblastic leukemia and contact with poultry in the first year of life (yes vs. no), restricted to children aged ≥1 year, meta‐analysis of 6 studies (1990–2013), Childhood Leukemia International Consortium.Click here for additional data file.Figure S5. Association between Acute Lymphoblastic Leukemia and Contact with Dogs in the First Year of Life (yes vs. no), Restricted to Children Aged ≥1 year, Meta‐Analysis of 12 studies (1980–2013), Childhood Leukemia International Consortium.Click here for additional data file.Figure S6. Association between acute lymphoblastic leukemia and contact with cats in the first year of life (yes vs. no), restricted to children aged ≥1 year, meta‐analysis of 12 studies (1980–2013), Childhood Leukemia International Consortium.Click here for additional data file.Table S1. Description of exposures of interest among controls, children aged 1 to 14 years old from all studies combined (1980–2013), Childhood Leukemia International Consortium.Table S2. Association between ALL and living in a farm, regular contact with animals in the first year of life, pooled stratified analyses of 11 studies (1980–2013) by ALL subtypes and by age, children aged 1 to14 years old, Childhood Leukemia International Consortium.Table S3. Association between ALL and living in a farm, regular contact with animals in the first year of life, pooled and one‐stage meta‐analyses of 13 studies (1980–2013), children aged 1 to 14 years old, Childhood Leukemia International Consortium.Table S4. Deterministic sensitivity analyses for home maternal pesticide use during pregancy as an Uncontrolled Confounder in the Investigations of the Association between Farm Residence, contact with livestock and pets in early childhood and ALL.Table S5. Deterministic sensitivity analyses for paternal smoking as an Uncontrolled Confounder in the Investigations of the Association between farm residence, contact with livestock and pets in early childhood and ALL.Click here for additional data file.Appendix S1. Acknowledgements and Funding by study.Click here for additional data file. Click here for additional data file.


H19 gene with the risk and prognosis of HCC. A total of 944 samples composed of 472 HCC patients and 472 matched controls were included in the risk analysis and amongst them 350 HCC samples were investigated in the prognosis analysis. KASP method was conducted for the SNP genotyping. The TT + CT genotype of rs2839698 was found to be associated with a 1.32-fold increased HCC risk  = 1.02–1.70). In the stratified analysis, rs2839698 (odds ratio (OR) = 1.57, P=0.007, 95% CI = 1.13–2.18) and rs3024270  were found to show more obvious increased HCC risk in the age ≤60 subgroup. And we found that rs2839698 showed an increased HCC risk in the ever smoking subgroup. But in the male subgroup of rs2735971, it showed a decreased HCC risk. Furthermore, haplotype analysis showed that rs2735971-rs2839698-rs3024270 G-T-C significantly increased the risk of HCC . Multilogistic analysis revealed no significant results of the interaction effects of the SNPs and environment factors. And in our study, rs2839698 showed a significant poor prognosis in the ever smoking subgroup (hazard rate (HR) = 5.19, 95% CI = 1.12–24.07, P=0.035). lncRNA-H19 rs2839698 SNP has the potential to be predictors for HCC risk and prognosis.Hepatocellular cancer (HCC) is one of the major causes of cancer-related mortality. Genetic polymorphisms may affect the susceptibility and clinical outcomes of cancers. We aim to manifest the association of single nucleotide polymorphisms (SNPs) of lncRNA- Hepatocellular cancer (HCC) is the major liver malignancy that attributes toward the second foremost cause of cancer-related mortality worldwide . IndividH19, located on chromosome 11p15.5 [H19 as an oncogene lncRNA in multiple cancers, such as, colorectal cancer [H19 plays important role in cancer initiation, progression, metastasis, and indicates poor prognosis and promotes tumor growth [In human genome, there are approximately 5–10% sequences transcribed constantly, and only approximately 1% are protein-coding sequences while a large part of others are non-coding RNAs (ncRNAs) . lncRNA,l cancer , breast l cancer , bladderl cancer , and so r growth .H19 works importantly in the incidence and prognosis of cancers and SNPs in H19 can be used as a promising biomarker for cancers risk [H19 polymorphisms and cancer risk had published [H19 SNPs and environmental factors as well as the association of H19 SNPs and the cancer prognosis were not analyzed further. And there is still no investigation about the H19 polymorphism associated with both HCC risk and prognosis. Whether lncRNA-H19 polymorphisms play some roles in HCC and could be promising biomarkers for the risk and prognosis of HCC, it is still not clear.It is well accepted that lncRNA-ers risk . Recentlublished , but theH19 gene according to the candidate gene association study strategy to explore the relationship between H19 polymorphism and HCC risk and prognosis. We aimed to manifest predictive biomarkers for risk and prognosis of HCC and provide the basic for the use of H19 gene polymorphisms as precautionary biomarkers of individuals and improve the comprehension of the etiology and disease development of HCC.In the present study, we selected three potential functional SNPs in lncRNA-This research project was approved by the Ethical Committee of the Shengjing Hospital of the China Medical University and written informed consent was obtained. The present study was designed as two independent but related parts including risk research and prognosis research. In the risk study, a total of 944 participants were involved, including 472 HCC patients and 472 sex and age (±5) frequency-matched controls from the Shengjing Hospital of China Medical University from 2013 to 2015. The response rate for cases and controls are up to 90% or more.H19 polymorphisms and overall survival in HCC patients, we conducted the research with the data of 350 HCC patients, whose information of death and survival was available for analysis. The HCC patients had pathologically confirmed HCC. Patients (i) with distant metastasis found preoperatively, (ii) who underwent preoperative radiotherapy or chemotherapy, or (iii) with incomplete pathological data entries were excluded from the prognosis analysis. Follow-up was completed by 10 July, 2017.For the aim to manifest the relationship between lncRNA-H19 were selected by the HapMap data [2 ≥ 0.8; a minor allele frequency (MAF) ≥5%; and Chinese Han Beijing (CHB) ethnicity. And we expanded 10 kbp both upstream and downstream of H19. Then, 17 SNPs were included as candidate SNPs , and we referred a published literature [The studied polymorphisms of lncRNA-−1 for genotyping. The genotyping assay was performed by Gene Company using KASP . The information of KASP primers was summarized in the Supplementary Table S1. Five percent of the whole samples were repeatedly genotyped, the concordance rate of the repeated cases performed 100% which suggested that the genotyping results were reliable.Genomic DNA was extracted by the method of literature  and was 2 test was used to compare the demographic characteristics of samples and ANOVA was conducted for age variability. Multivariate logistic regression with adjustments for age and gender was proceeded to calculate the association of the selected SNPs and HCC risk. SHEsis software was used to analyze the haplotype of the selected gene [2 test. Univariate and multivariate survival analysis was conducted by the log-rank test and the Cox proportional hazards model. Statistical analysis was performed by using SPSS version 18.0 software  and P-value <0.05 was considered to be significantly statistical.χted gene . The anaH19 SNPs  which were all conformed to Hardy–Weinberg equilibrium (HWE).The demographic characteristics of HCC and controls are shown in Supplementary Table S2. In H19 rs2839698 polymorphism was calculated to be associated with an increased risk of HCC. In the dominant model, rs2839698 TT + TC genotype appeared with a 1.32-fold increased HCC risk when compared with CC genotype  and the similar situation was showed in rs3024270 CC genotype of age ≤60 subgroup . When stratified by smoking factor, rs2839698 showed a more obvious HCC tendency in CT subgroup of ever smoker subgroup  = 1.01–1.51, P=0.043)  .H19 polymorphisms and environmental factors in HCC risk . These results suggested that the G-T-C haplotype suffers more risk than other haplotype.Furthermore, we performed interaction analysis for the multiple lncRNA-H19 SNPs and HCC prognosis. No significant association was found between H19 SNPs and HCC overall survival. In addition, we performed subgroup analysis for HCC prognosis and found rs2839698 that showed significant poor survival condition in the ever-smoking subgroup which suggested that it may affect HCC prognosis and could be a promising biomarker for HCC prognosis. As discussed above, the expression of H19 could be induced by cigarette smoke [We further performed univariate and multivariate Cox proportional hazards regression analysis of overall survival time to explore the association between lncRNA-te smoke . Thus, wHowever, there still existed several limitations in the present study. First, the sample size was relatively not large enough for the analysis which may limit the possible analysis of other subgroup analysis and interaction analysis for variant genotype. Second, we only studied the local population and did not include residents of other areas. Third, because the controls were collected from the health check program of our hospital and there was no message of the HBV for them which could not assess the influence of this factor. In future, larger sample and multicenter samples are needed for the confirmation study of our findings.H19 was associated with an increased risk of HCC. And more significant findings were shown in the age ≤60 subgroup in rs2839698 and rs3024270. In the ever-smoking subgroup, rs2839698 showed an obvious increased HCC risk too. But we got an adverse result in the male subgroup of rs2735971 SNP that showed a decreased HCC risk. In addition, we found that the rs2735971-rs2839698- rs3024270 G-T-C significantly increased the risk of HCC in the analysis of haplotype and HCC risk. And the MDR analysis had no significant findings. In the prognosis analysis, the rs2839698 showed a poor prognosis in the ever-smoking subgroup. In the future, the larger scale sample experiments and analyses are needed to confirm our results.In conclusion, we found an intron rs2839698 SNP of lncRNA-