Background: Increased consumption of fruit and vegetables has been shown to be associated with a reduced risk of cognitive impairment and dementia in many epidemiological studies. The purpose of this study was to assess the strength of this association in a meta-analysis.Methods: We identified relevant studies by searching Medline, Embase, and Cochrane Library electronic databases (from 1970 to January 2016). Study were included if they reported relative risks and corresponding 95% confidence intervals (CIs) of cognitive impairment and dementia with respect to frequency of fruit and vegetable intake.Results: Nine studies  met the inclusion criteria and were included in the meta-analysis. There were a total of 31,104 participants and 4,583 incident cases of cognitive impairment and dementia. The meta-analysis showed that an increased consumption of fruit and vegetables was associated with a significant reduction in the risk of cognitive impairment and dementia . Subgroup analysis indicated this inverse association was only found among participants with mean age over 65 years and combined sexes. Dose–response meta-analysis showed that an increment of 100 g per day of fruit and vegetable consumption was related to an approximately 13%  reduction in cognitive impairment and dementia risk. There was no potential publication bias in the meta-analysis and the dose–response meta-analysis.Conclusion: The increased consumption of fruit and vegetables is associated with a reduced risk of cognitive impairment and dementia. Cognitive impairment is an important public health concern, with a prevalence that is expected to rise with the population aging. It is a transitional stage between normal aging and dementia. Interventions may be particularly effective at earlier stages of disease development, which offer an opportunity for reducing the public health burden of Alzheimer’s disease (AD) and other dementias through early detection and prevention .A number of epidemiologic researches and animal models have found associations between dietary components and age-related cognitive impairment and dementia , and adhWe followed the guidelines published by the Meta-analysis of Observational Studies in Epidemiology (MOOSE) group to complete the meta-analysis (Supplementary Table A study was eligible for inclusion if the following criteria were met: (1) examination of dietary consumption of fruit and/or vegetables as the variable of interest; (2) determination of incidence of cognitive impairment or dementia as the outcome of interest; and (3) reporting the relative risks of cognitive impairment or dementia calculated according to the highest category with the lowest category of fruit and/or vegetables consumption, and their 95% confidence intervals (CIs). The studies about animal experiment, mechanistic research, and review research were excluded.Two researchers independently extracted the following data from each publication: author, country, study design, sample size, disease type (cognitive impairment or dementia), number of cases, age, disease ascertainment, exposure variable , exposure assessment, risk estimates with CIs, and factors adjusted for. The most adjusted estimate was included when a study reported more than one risk estimate. The quality of each study was assessed by two researchers, using the Newcastle-Ottawa Scale recommended by We performed meta-analyses of risk estimates for cognitive impairment and dementia comparing the highest category of exposure of fruit and vegetables with the lowest category. We pooled the data on cognitive impairment and dementia together, since there were limited studies on each single disease. Dose–response meta-analyses of fruit and vegetable consumption and risk of cognitive impairment and dementia were then conducted using methods previously reported , which fI-squared (I2) statistic and Q-statistic were used to explore the heterogeneity among studies. Large I2 (>50%) or P < 0.10 for Q-statistic suggests substantial heterogeneity among studies. We also performed a sensitivity analysis by removing each individual study from the meta-analysis . A significant heterogeneity was observed for the pooled analyses . Subgroup analyses showed that the inverse association between fruit and vegetable consumption and risk of cognitive impairment and dementia was only found in the participants with mean age over 65 years  and combined sexes , and studies from the United States and with full marks of study quality score did not show statistical significances (Table 2). Sensitivity analysis indicated that the inverse association was not materially changed in the leave-one-out analyses by omitting one study in turn, with a pooled OR of cognitive impairment and dementia range from 0.74 (95% CI 0.69–0.80) to 0.83 (95% CI 0.76–0.90) for the highest vs. lowest category of fruit and vegetable consumption .Figure 4). There was no potential heterogeneity among studies .Only four estimates from three individual studies were included in the dose–response meta-analysis, because there were only two categories of fruit and vegetable consumption in other studies, and dose–response meta-analysis requires data for the distribution of cases and person-time across at least three categories of exposure . In the Figure 5) showed that the studies were distributed fairly symmetrically about the combined effect size in both the meta-analysis and the dose–response meta-analysis, which suggests little publication bias. Egger’s regression test  and Begg–Mazumdar test  also showed that there was no potential publication bias in the meta-analysis and the dose–response meta-analysis.Visual assessment of funnel plots . But in our meta-analysis, most included studies used fruit and vegetable combined as the exposure variable, and there were only three studies included in the subtotal estimates of both fruit alone and vegetable alone, leading to a low statistical power. This is also the reason why we pooled the results of fruit alone, vegetable alone, and fruit and vegetable combined together. More studies are needed to further assess the role of fruit and vegetable separately on the risk of cognitive impairment and dementia. In addition, our meta-analysis restricted to overall fruit and vegetable consumption, some particular kinds of fruit and vegetable intake, such as nuts, cruciferous vegetables, dark and green leafy vegetables, root vegetables, cabbage, tomatoes, soybeans, and soybean products . This means that the protective effects of fruit and vegetables against cognitive impairment and dementia are stronger for Chinese populations than the Western populations. This may be explained by the different dietary patterns and food preparation methods between Chinese and Western populations. Firstly, the proportion of saturated fat-rich food is higher in the Western dietary pattern, which may increase the risk of cognitive impairment and dementia as discussed above. Secondly, in Western world, people mainly eat raw vegetables, such as in the form of salad, while the Chinese populations often eat cooked, steamed, or boiled vegetables. Usually, people tend to eat more vegetables after being cooked, and the cooked vegetables can be more easily digested because the cell wall of raw vegetables is relatively harder, which will increase the burden of digestion. In addition, studies show that the nutritional quality increases in all cooked vegetables because of matrix softening and increased extractability of compounds, which could be partially converted into more antioxidant chemical species .The strengths of the present meta-analysis include the considerable number of studies and subjects included, as well as the acceptable methodologic quality of the studies on which the analysis is based. The study has several limitations. Firstly, a substantial heterogeneity across studies was apparent in the meta-analysis. The heterogeneity was not accounted by age, sex, geographic location, study design, disease type, and dietary assessment method. However, the subgroup analysis for study quality score reduced the heterogeneity, which means the study quality is a main source of heterogeneity. Secondly, the assessment for fruit and vegetable consumption is mostly based on self-reported habits, and such data are subject to recall errors. In addition, there were many other differences among studies, including dietary assessment methods, the variety of fruit or vegetables investigated, the definition of the reference group, and the choice of exposure categories. These differences could affect the estimation of the true relation. Thirdly, the meta-analysis is based on observational studies, which leaves the possibility that potential confounders cannot be ruled out, affect the relation between fruit and vegetable consumption and risk of cognitive impairment and dementia. A meta-analysis is not able to solve problems with confounding that may be inherent in the included studies. However, most studies have made adjustment of major confounding factors, which should reduce the potential bias due to the dietary and lifestyle factors. Finally, due to the limited studies, we combined consumption of fruit and vegetable together as the exposure variable and cognitive impairment and dementia together as the outcome. The effects of fruit and vegetable separately on cognitive impairment or dementia should be further investigated in future studies.This meta-analysis indicates significant inverse association between fruit and vegetable consumption and risk of cognitive impairment and dementia. The risk of cognitive impairment and dementia was reduced by 20% for a higher consumption of fruit and vegetables, and by 13% for an increment of 100 g per day of fruit and vegetable consumption. For further studies, based on our findings, we suggest that the investigators should improve the standardization of various dietary assessment methods, which may make the results more accurate and conceivable. Furthermore, the use of genetic and biological makers as surrogate end points in the future studies should help to clarify the cause and effect relationship that link fruit and vegetable consumption and cognitive impairment and dementia.ZZ designed the study. JH, DS, and RD collected the data. XJ performed all analyses. XJ, JH, JW, and ZZ wrote the manuscript. All authors contributed to writing of this manuscript.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.


Fish, rich in ω-3 polyunsaturated fatty acids, has been found to be associated with lower risk of several types of cancer risk, and beneficial for brain development. However, the association between fish intake and brain tumor risk is still inconsistent. Therefore, we conducted a meta-analysis to clarify the association.Relevant studies were identified from PubMed and EMBASE databases. The pooled relative risks were obtained by the fixed-effects model when no substantial heterogeneity was observed. Otherwise, the random-effects model was employed. Subgroup and publication bias analyses were also performed.Nine observational studies were included in the meta-analysis. The pooled relative risk of brain cancer for the highest vs. lowest category of fish intake was 0.83 . No significant heterogeneity was detected. Dose-response analysis showed that the RR per 100 g/day increase in fish intake was 0.95 (95% CI: 0.91–0.98). The results remained unchanged in subgroup and sensitivity analyses.The results of our meta-analysis suggest that fish intake might be associated with lower risk of brain cancer risk. The finding should be further confirmed by future cohort studies with validated questionnaires and strict control of confounders.The online version of this article (doi:10.1186/s12937-016-0223-4) contains supplementary material, which is available to authorized users. Brain tumor is the neoplasmsin primary central nervous system, the incidence of brain cancer is approximately 14.4 per 100,000 persons annually, among which gliomas andmeningiomas are the two most common types, accounting for more than 80% cases , 3.Apart from ionizing radiation and some genetic mutations, the risk factors of brain cancer are still not well understood to date . LifestyAs one important component of diet, fish is considered to be quite healthy, which has been found to be associated with lower risk of all-cause mortality , cardiovWe performed and reported our meta-analysis, according to the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines . RelevanStudies were included in our meta-analysis if they met the following criteria: 1) observational studies with cohort or case-control design; 2) the exposure of interest was fish intake, total fish or fresh fish intake, consumption of smoked, pickled or salted fish was not included; 3) the outcome was the risk of brain cancer , including both incidence and mortality; 4) relative risks[RR] (or odds rations [OR] or hazard ratios [HR]) and their corresponding 95% CIs were reported. We did not include the studies exploring the effect of fish intake on the survival of patients with brain tumor. If multiple articles were reported from the same population, then the latest one with the most information was included.Two authors independently reviewed the identified articles. The following information was extracted: the first author’s name, study design, location, publication year, study period, comparisons, RRs with their 95% CIs, adjusted confounding factors. The quality of each original study was also evaluated by the Newcastle-Ottawa scale, which is widely employed for assessing the observational studies . BrieflyI2 statistics [p < 0.1 was supposed statistically significant. In this case, the random-effect model was used for pooling the RRs. Otherwise, the fixed-effects model was employed [RRs with their 95% CIs adjusting for most confounders in original studies were extracted for data synthesis. In some studies, RRs were reported stratified by gender or type of fish, we combined them using a fixed-effect model. Considering the differences of exposure categories in each original study, RRs comparing the highest with the lowest categories of fish intake were employed to gain a pooled estimate. Heterogeneity was evaluated using the Q andatistics . For theemployed .We also attempted to perform a dose-response analysis, according to the method proposed by Greenland and Longnecker  and OrsiP < 0.1), we used the “trim and fill” method to correct the results [In addition, we also conducted subgroup analyses by possible confounders, and meta-regression analyses to investigate the sources of heterogeneity. Influence analysis was carried out in which one individual study was removed each time, to explore the effect of each study on the overall result. Publication bias was assessed by funnel plot and Egger’s test , if ther results . All staOriginally, 1708 records were found in the electronic databases, 1679 were excluded by title and abstract review, and 29 articles were further checked by full-text, a total of nine studies were included in our meta-analysis finally. The flow diagram of literature search process was shown in Fig. Overall, the quality scores of included studies were not quite high, with an mean of 5.78, three studies scored five points , 23, 25,I2 = 37.5%, P = 0.12. Thus, a random-effects model was used to combine the RRs, we observed a significant inverse association between fish intake and the risk of brain tumor, with a pooled RR of 0.83 (95% CI: 0.70–0.99), as shown in Fig. A moderate heterogeneity was observed among studies, with I2 = 51.7%; Pheterogeneity = 0.04) , so a linear model was used to evaluate the dose-response relationship between fish intake and brain tumor risk, as shown in Fig. All the original studies reported the results for at least three quantitative levels of fish intake, and were included into the dose-response analysis. The results showed that the summary RR per 100 g/week increment in fish intake was 0.95 (95% CI: 0.91–0.98), with an evident heterogeneity 04) Fig. . The P v Fig. . TI2 = 8.2%; Pheterogeneity = 0.36), and the RR for two hospital-based case-control studies was 0.62 (95% CI: 0.26–1.47), with significant heterogeneity , however, only one cohort study was included in the meta-analysis, and the result was non-significant [In subgroup analyses stratified by study design, the pooled RR for five population-based case-control studies was 0.83 (95% CI: 0.72–0.96) using a fixed-effects model, with little heterogeneity . The results for the association between meningioma and fish intake were always presented in combination with other brain tumors and ineligible for subgroup analysis. Besides, Terry MB et al. [After stratifying by pathology types, five studies , 26, 28 P = 0.74), study design (P = 0.26) location (P = 0.39) and number of cases (P = 0.75). After removing the study by Hu J et al. in 1999 year [I2 of 51.7% to 13.0%. Influence analysis by omitting one study each time showed that the overall results were not changed significantly, ranging from 0.79 (95% CI: 0.64–0.99) to 0.88 (95% CI: 0.77–1.00).Meta-regression analyses were performed to explore the sources of heterogeneity, however, none of the confounding factors were found to be significantly associated with heterogeneity, including publication year . The “trim and fill” method was used to correct such bias, however, no studies were found to be filled, and data remained unaltered, implying that the effect of publication bias on the overall results might be tiny.The funnel plot seemed to a little unsymmetrical Fig. , and EggTo our best knowledge, this is the first meta-analysis with systematic review on the association between fish intake and brain cancer risk. Based on nine observational studies, involving a total of 4428 brain cancer cases and 501,617 participants, the results showed that fish intake might be associated with lower brain cancer risk, and dose-response analysis also supported the conclusion. However, due to the small number of studies included, the statistical power was weakened to some extent, and further subgroup analyses were limited.Although the precise mechanisms whereby fish intake reduces the risk of brain cancer are still unknown, the biological explanation is plausible. Fish is rich in polyunsaturated fatty acids (PUFAs), high quality proteins and minerals, and low in fat and cholesterol, especially, marine fish contains plenty of omega-3 fatty acids, including docosahexaenoic acid (DHA) and eicosapentaenoic acid (EPA), which exert a series of health benefits and neuroprotective properties , 41. StuNotably, our analysis is focused on total fish intake, including fresh and frozen fish, however, smoked, pickled or salted fish was not included in our study, further literature review found that five studies –24, 47 iThe strength of our study lies in a large sample size  and long duration (from 1986 to 2006), thus to enhance the statistical power to detect possible weak association. Moreover, dose-response analysis was conducted to quantify the association. However, several shortcomings must be also acknowledged in our study when interpreting the results. Firstly, the original studies included were mostly low-quality case-control studies, which were prone to recall and selection bias, and there was only one prospective cohort study with a null finding between fish intake and brain tumor risk. Secondly, unvalidated questionnaires and inadequate adjustment for potential confounders, including irradiation exposure, total energy intake and body mass index, would also influence the stability of the results. Thirdly, characteristics of study population, sample size, follow-up duration, and exposure levels of fish intake were different across the original studies, which might contribute the heterogeneity among studies and lead to incomparability of results, however, both highest category of fish intake vs. the lowest analysis and dose-response analysis were conducted and the results were consistent. Besides, since the number of included studies was small, subgroup analyses by common confounders including location, design, publication year and types of fish were not performed. Lastly, we only searched publications in English, possible publication bias was inevitable. Indeed, there was an indication of publication bias with Egger’s test. However, the statistical power of the test was relatively low due to the small number of studies included, and no studies were added when employing the “trim and fill” method, suggesting that the influence of publication bias on results was insignificant.In summary, the present meta-analysis suggests that fish intake is associated with lower risk of brain tumor risk. Given the small number of included studies, the finding should be further confirmed by further prospective cohort studies with validated questionnaires and strict control of confounders.


Obesity is found to increase the risk of most cancer types, but reduce lung cancer risk in many studies. However, the association between obesity and lung cancer is still controversial, mainly owing to the confounding effect of smoking.Eligible studies were identified from electric databases to July 1, 2017. Relevant data were extracted and pooled using random-effects models; dose-response and subgroup analyses were also performed.P < 0.01) for excess body weight (body mass index [BMI] ≥ 25 kg/m2). An inverse linear dose-response relationship was observed between BMI and lung cancer risk in never smokers, with an RR of 0.89 per 5 kg/m2 increment in BMI. The results remained stable in most subgroup analyses. However, when stratified by sex, a significant inverse association existed in women but not in men. Similar results were found in analyses for other categories of BMI.Twenty-nine studies with more than 10,000 lung cancer cases in15 million never smokers were included. Compared with normal weight, the summary relative risk (RR) was 0.77 contains supplementary material, which is available to authorized users. Obesity is one of the most important risk factors for several major non-communicable diseases, including cardiovascular diseases, diabetes, and cancer, and the widespread prevalence of obesity is becoming a major threat to global public health . AccumulAs one of the most common cancers in both men and women, lung cancer causes more deaths than any other cancer . Curious. also reported an attenuated linear dose-response association between BMI and lung cancer risk  in non-smokers, without statistical significance [Lung cancer in never-smokers accounts for approximately 10–15% of all lung cancer patients and causes more than 15,000 deaths annually . Concernificance . In the obesity, overweight, body mass index, body size, leanness, or anthropometric in combination with lung cancer, lung carcinoma, or lung neoplasm. Our literature search was restricted to the full-text publications, and no language restriction was applied. The reference lists of identified articles and other similar meta-analyses were also checked to find additional studies.We searched the PubMed database to find relevant studies from January 1, 1966, to July 1, 2017.The following key words were used: 2), including the categories of obesity, overweight, underweight or excess weight; 3) relative risk (RR) estimates (or hazard ratios or odds ratios) and 95% confidence intervals (CIs) for never smokers were reported or could be calculated from the data. 4) the outcome was the incidence or mortality of lung cancer; 5) observational studies with a cohort or case-control design. When duplicated studies were reported from the same population, the ones with the longest follow-up were included.Two independent investigators reviewed all the records and included studies that met the following criteria: 1) study population was never (or non-) smokers, current and former (past or ex-) smokers were not considered in this study, never smokers are defined as those who have not smoked greater than 100 cigarettes in their lifetimes and do not currently smoke; 2) the exposure of interest was BMI . A random-effects model was used to pool the individual RRs, considering the heterogeneity among studies, which was evaluated by the Q and I2 statistics [Obesity and overweight were defined as BMI ≥30 and 25–29.99 kg/mg et al. . A fixedatistics .2increment in BMI and pooled together using a random-effects model. To compute the study-specific slope from the correlated log RR estimates across BMI levels, a two-stage generalized least-squares method with fractional-polynomial regression models was employed [Only studies that reported RRs with 95% CIs for at least 3categories were included into the dose-response analysis using the method proposed by Greenland  and Orsiemployed . To testp < 0.10 was regarded as statistically significant, and the trim-and-fill method was used to adjust for potential bias. All statistical analyses were done with the STATA version 12.0 software .We also carried out subgroup analyses stratified by potential confounding factors, including study design, outcome, sex, diagnosis method, ethnicity, and quality. Meta-regression analyses were performed to explore the sources of heterogeneity. To evaluate the stability of the results, sensitivity analysis was employed to examine the change of pooled results after removing one study each time. Publication bias was assessed by funnel plot and Egger’s test, In total, 3937 articles were identified from the databases, and after removing the ineligible studies, 29 studies were included in the meta-analysis, including 21 cohort studies and 8 case-control studies . Among all studies, 21 were considered as high quality (≥7 points). The most common confounders adjusted in original studies included age, sex, alcohol consumption, vegetable/fruit intake, and physical activity; however, few studies were controlled for total calorie intake, other chronic diseases, concomitant medication, or environmental status. The baseline characteristics of all studies are shown in Additional file P = 0.01) and 0.76 compared with the normal category. Combined analysis of 23 studies showed that the RR was 0.77 for the excess weight category  for the excess weight category.Subgroup analyses suggested that RRs did not differ significantly by design, outcome, cancer ascertainment, BMI assessment, quality, or whether important confounders were adjusted for in original studies, although some results were statistically negative, mainly owing to the small number of studies included. When stratified by sex, some differences were observed; the results for women were consistently significant for all categories, whereas no positive associations were found for men Table . To avoiP < 0.01) per 5 kg/m2 increase in BMI, with a high heterogeneity (I2 = 86.44%)  for cohort studies and 0.90 for case-control studies (Table P < 0.01) and 0.96 for women and men, respectively  per 5 kg/m2 increase in BMI.Finally, 28 studies were included in the dose-response analysis; the summary RR was 0.89 Fig. .The RRs p for nonlinearity = 0.18), and an inverse linear trend was fitted in a random-effects meta-regression model , 0.87, 0.81, and 0.80  for BMI of 15, 25, 30, and 35 kg/m2, respectively. When stratified by sex, the inverse linear trend was still present for women but disappeared for men.No evidence of a nonlinear relationship between BMI and lung cancer risk in never smokers was found , but no studies were needed to be filled with the use of trim and fill method, suggesting that the influence could be negligible. In fact, the bias might be caused by insufficient data reported in original studies on the category of obesity, since in other category analyses, the funnel plots seemed to be symmetrical, and no significant publication biases were found.As described above, substantial heterogeneity was observed across studies, but meta-regression analyses showed that most of the confounders including BMI assessment, ethnicity, design, quality, outcome, and diagnosis method were not significantly associated with the heterogeneity. After excluding the 2 outlier studies by Kabat et, al.  and Kond[et, al. , the hetIn the pooled analysis of 29 observational studies, involving more than 10,000 lung cancer cases in 15 million never smokers, the results suggested that higher BMI was associated with lower lung cancer risk, especially in women. In contrast with previous meta-analyses, our study includes the largest sample up to now, and the results were stable both in the subgroup and dose-response analyses.Previous studies reported that obesity was associated with a lower risk of certain cancer types, particularly smoking related-cancers , 12, 31.As we mentioned previously, several hypotheses have been put forward to explain the inverse association between lung cancer risk and BMI. As our study was limited to never smokers, the confounding of smoking, one of the most common arguments for the trend, was avoided as much as possible. A second hypothesis is that the lower BMI might reflect preclinical weight loss caused by early lung cancer itself or other related diseases. To solve this doubt, only studies in which BMI was measured5 or more years before diagnosis were analyzed, and the inverse association remained unchanged. In fact, the participants with previous clinical weight loss were excluded at recruitment in most studies. It was also speculated that socioeconomic factors might be relevant to the inverse association, such as indoor air pollution in developing countries; however, our results indicated that the inverse association was stable across the strata of ethnicity.et al. found that BMI was more strongly related to lower lung cancer risk in women than in men in a large cohort study [Interestedly, we found a sex difference in the association between BMI and lung cancer risk, although p-interaction for sex is not statistically significant. Previous studies reported paradoxical results concerning the differences between sexes , 12, Smirt study . The autrt study , which w+ T cells to maintain normal immune functions [Not only epidemiological studies revealed the possibility that obesity might lower lung cancer risk, but some biological discoveries also provided useful hints. It has long been noted that lower BMI might increase the susceptibility of DNA to chemical carcinogens in cigarettes , and excunctions . In addiunctions . Paradoxunctions , is alsounctions . These sThe results of our study should be interpreted with caution, as associations found in a meta-analysis of observational studies do not reveal causation. Several other limitations should also be considered. First, concerning the outcomes of interest, both the incidence and mortality of lung cancer were included. It is reasonable to do this since lung cancer is relatively rare in the overall population, and previous studies reported that the incidence and mortality of lung cancer almost coincided with each other . Second,Every sword has two edges. Obesity has been stereotyped as a risk factor for many chronic diseases including most types of cancer, but our study shows that higher BMI is associated with lower lung cancer risk, especially in women. Our results do not suggest increasing body weight to decrease the risk of lung cancer; however, underweight is also inadvisable, and maintaining a proper weight is the best choice. The results of our study are helpful to explain the J-shaped association between BMI and total mortality . FurtherIn conclusion, the results of our meta-analysis indicate that higher BMI is associated with lower lung cancer risk, especially in women, which alter our common understanding of the relationship between obesity and cancer, although the causal relationship between these two factors cannot be determined from this analysis. Additional studies are required to validate these findings and to better understand the biologic rationale for this observation.Additional file 1:Table S1. Characteristics of studies included in the meta-analysis of obesity and lung cancer risk in non-smokers. (DOCX 88 kb)Additional file 2:Table S2. Quality scores of the cohort studies included in the meta-analysis, assessed by the Newcastle-Ottawa scale. (DOCX 20 kb)Additional file 3:Table S3. Quality scores of the case-control studies included in the meta-analysis, assessed by the Newcastle-Ottawa scale. (DOCX 18 kb)


However, the beneficial effect of coffee consumption on thyroid cancer was found only in hospital-based case-control studies . There was no significant association between coffee consumption and thyroid cancer risk according to our meta-analysis results. These findings should be interpreted with caution because of potential biases and confounding variables. Further prospective studies with a larger number of cases are encouraged to confirm these results.An inverse association has been reported between coffee consumption and the risk of several cancers. However, the association between coffee and thyroid cancer is controversial. Thus, this study aimed to evaluate the association between coffee consumption and the risk of thyroid cancer through a systematic review and meta-analysis. Published studies were examined from PubMed, Embase, Cochrane Central, and the reference lists of the retrieved articles. The summary odds ratio (OR) for the association between coffee consumption was categorized as highest versus lowest consumption, and thyroid cancer risk was calculated using a fixed effects model. Subgroup analyses by study design, geographic location, source of controls, and adjusted variables were performed. A total of 1039 thyroid cancer cases and 220,816 controls were identified from five case-control studies and two cohort studies. The summary OR for the association between coffee consumption and thyroid cancer risk was 0.88  = 0.71–1.07). There was no significant heterogeneity among the study results (I² = 0%,  Thyroid cancer is the eighth and 18th most common cancer among women and men, respectively, in the world . SeveralCoffee is one of the most widely consumed beverages in the world . Coffee Although many studies have analyzed coffee consumption and its effects on cancers, studies assessing the association between coffee consumption and thyroid cancer risk have yielded inconsistent results. For instance, a case-control study including 284 thyroid cancer cases showed that thyroid cancer patients drank less coffee than healthy patients . HoweverTo our knowledge, a meta-analysis of recently published data has not been performed. Thus, in this study, we performed a meta-analysis to determine the association between coffee consumption and the risk of thyroid cancer. We investigated the risk of thyroid cancer for the highest coffee consumption compared with the lowest consumption.A comprehensive systematic search was performed for relevant studies up to 24 June 2016 in PubMed, Embase, and Cochrane Central. Medical Subject Headings, Emtree headings, and other relevant keywords were used to perform the literature search. The keywords used for the search were (caffeine OR coffee) AND (thyroid) AND (tumor OR tumor OR cancer OR neoplasms). In addition, the references from relevant research and review articles were screened to identify additional studies. This review was conducted in line with the preferred reporting items for systematic reviews and meta-analysis (PRISMA) statement .Two investigators (Mi Ah Han and Jin Hwa Kim) independently reviewed the studies. If there was disagreement, consensus was obtained by discussion. The titles and abstracts were reviewed to identify potentially relevant studies. Full-text manuscripts were reviewed if they met all the following inclusion criteria: (a) studies were based on humans; (b) information on coffee consumption was provided; (c) studies were focused on thyroid cancer; (d) information on the association between coffee consumption and thyroid cancer was provided, including estimates of or calculable information for the OR, risk ratio, or hazard ratio and their corresponding CI; and (e) full-length papers in English since 1990. Reviews and pooled analyses articles were excluded. Since 1990 the coffee market has been subject to the free market forces of supply and demand  and thenTwo reviewers (Mi Ah Han and Jin Hwa Kim) used a predesigned data abstraction form to independently collect data from eligible studies. From the selected studies, the following data were extracted: first author’s last name, publication year, country, study design, number of cases, age range, gender, coffee type (caffeinated versus decaffeinated), coffee consumption frequency, type of thyroid cancer, study-specific estimates, and adjusted variables.2 statistic above 50%. In addition, several subgroup analyses were performed to assess the robustness of the summary estimates. The subgroups were analyzed by study design, geographic location, source of controls, and adjusted variables. Publication bias was not assessed as there were inadequate numbers of included studies to properly assess a funnel plot. Statistical analyses were performed using Reviewer Manager version 5.3 . Statistical significance was defined as a p-value < 0.05.An overall summary OR and its 95% CI for highest coffee consumption compared with lowest  consumption was estimated based on the individual estimates from each study and compared across studies using a forest plot. Each study estimate was given a weight based on the inverse of the variance of the effect estimate. A fixed effect model was used to summarize the study-specific estimates. However, a random effects model was used if there was heterogeneity among study results, which was considered as an IA total of 107 articles published before 24 June 2016 were identified from PubMed, Embase, and Cochrane Central. Six additional articles were retrieved from the reference lists of the identified publications. From these 113 articles, 22 duplicates were removed. The titles and abstracts of the remaining 91 articles were reviewed for relevance. After screening, 16 articles were selected and assessed for eligibility. From these 16 articles, nine were excluded due to the following reasons: one study did not consider thyroid cancer as an outcome, two studies did not assess coffee as an exposure, one study did not provide calculable information for the OR, four studies were review articles, and one case-control study involved thyroid cancer patients for both the case group (anaplastic thyroid cancer) and the control group (papillary cancer). Therefore, seven articles were included in the meta-analysis ,22,23,24A total of 221,855 participants (1039 thyroid cancer cases) from the seven studies (two cohort studies and five case-control studies) were included in the meta-analysis. From the seven studies, two studies were conducted in the USA, one in Greece, two in Japan, one in Germany, and one in Norway and Sweden. Three studies provided information on caffeinated coffee intake, and five studies provided adjusted risk estimates .p = 0.79, I2 = 0%) (Nine comparisons from the seven studies showed a link between coffee consumption and thyroid cancer risk. The summary OR for the association between thyroid cancer risk and highest versus lowest coffee consumption was 0.88 (95% CI = 0.71–1.07). In addition, the heterogeneity among the included studies was not statistically significant (I2 = 0%) .p ≥ 0.05). In the subgroup analysis by the source of controls, an association between coffee consumption and a reduced risk of thyroid cancer was found only in the hospital-based case-control studies  but not in the population-based case-control studies . Among hospital-based case-control studies, the summary OR was 0.59 (95% CI = 0.37–0.93). However, hospital-based controls might alter their coffee consumption behavior according to their medical condition . When coCoffee contains a variety of biologically active compounds that might affect the risk of cancer. For instance, caffeine increases intracellular cyclic adenosine monophosphate, which has an inhibitory effect on cell  growth ,27. HoweMost study-specific estimates included in this meta-analysis were adjusted for several risk factors for thyroid cancer such as age ,20,24, sThe coffee consumption rate, awareness about the components of coffee and its effect on health and body were significantly different by gender . This diThere are limitations of this meta-analysis that call for caution when interpreting the results. First, the majority of included studies were retrospective, whereas only two studies were prospective cohort studies. Retrospective studies are prone to recall bias . AlthougTo our knowledge, this is the first meta-analysis to evaluate the association between coffee consumption and thyroid cancer risk. Our meta-analysis includes a large number of thyroid cases, which enables a greater possibility of reaching reasonable conclusions about the link between coffee consumption and thyroid cancer risk.The results of this meta-analysis suggest that coffee consumption has no effect on the risk of thyroid cancer with the exception hospital-based case-control studies. These findings should be interpreted with caution because of potential biases and confounding variables. Future studies are suggested to assess the difference by gender and histological type of thyroid cancer. Also, prospective studies that consider more detailed coffee consumption behavior  and the dose-response relationship between coffee consumption and thyroid cancer risk are necessary to support our results.


Background. Previous studies have revealed that gene polymorphisms of inflammatory factors may influence the development or progression of periodontitis, a main cause of tooth loss in adults; however, due to limitations of individual studies, inconsistent findings were reported. Objective. To meta-analytically investigate the relationship between periodontitis and the Interleukin-4 (IL-4) and Interleukin-4 receptor (IL-4R) gene polymorphisms. Methods. Databases were searched for relevant case-control studies. After study selection based on the predefined selection criteria, methodological quality assessment and data extraction were conducted independently by two reviewers, before subsequent statistical analyses. Results. 37 studies involving 4,385 patients and 5,168 controls were included. All the studied IL-4 polymorphisms were not significantly associated with periodontitis, except the -33C/T  associated with reduced AgP susceptibility. Positive association was found between IL-4R Q551 polymorphism and periodontitis susceptibility in three genetic models . Conclusions. A positive association exists between the IL-4R Q551R polymorphism and occurrence of CP. The IL-4 -33 CT genotype is negatively associated with the occurrence of AgP. Among tγc chain .Currently, studies investigating the association between IL-4 gene polymorphisms and periodontitis mainly focus on its promoter region. Among the polymorphisms studied, single nucleotide polymorphisms (SNP), including the -590C/T (rs2243250), -33C/T rs2070874), -1099T/G (rs2243248), and the -70-bp variable number of tandem repeats (VNTR), are the most frequently reported , 7. The 0874, -10The reporting of the present study follows the Meta-analysis Of Observational Studies in Epidemiology (MOOSE) statement .Databases including the PubMed, Embase, Scopus, ScienceDirect, Web of Science, CBM, CNKI, and WanFang were searched to identify relevant published papers addressing the association between IL-4 or IL-4R gene polymorphisms and periodontitis, up to August 1, 2016. The following search terms were used: “interleukin-4”, “interleukin 4”, “IL-4”, “Interleukin-4 receptor”, “IL-4R”, “periodontitis”, “periodontal disease”, and “polymorphism”. Full-text publications and their reference lists were carefully screened to decide whether information on the topic of interest was included. Additionally, the search was expanded by reviewing special meeting issues of journals in order to retrieve relevant abstracts. Studies that met the following criteria would be included: (i) case-control study on the IL-4 and IL-4R gene polymorphisms and susceptibility to periodontitis including chronic periodontitis (CP) and/or aggressive periodontitis (AgP); (ii) research subjects being patients with periodontitis and healthy controls; (iii) Reported data adequate for estimating the odds ratio (OR) with 95% confidence interval (95% CI); (iv) being published in English, Chinese, or Russian. On the other hand, studies with incomplete data and pedigree analysis and duplicated reports of the same study were excluded.Two investigators (Jia and Yuan) independently extracted the following data from each included study: first author's surname, publication year, country, ethnicity, type of disease, source of control, sample size, percentage of smokers among patients, genotyping method, genotype distribution in cases and controls, and Hardy-Weinberg equilibrium (HWE) for controls .The included studies were evaluated in terms of methodological quality using the Newcastle-Ottawa scale (NOS) by two authors (Jia and Yuan) independently. Any discrepancy between the two authors was solved by discussion with a third investigator (Zeng).I-squared statistic and assessed in terms of significance using the Chi-square based Q-test. I2 > 50% and/or P < 0.1 indicated significant heterogeneity among studies, in which case the random-effects model was used to perform the meta-analysis; otherwise, the fixed-effects model was used. Subgroup analyses were conducted with stratification by HWE, ethnicity, smoking status, and periodontitis type. Sensitivity analysis was performed to examine the robustness of the results. The potential publication bias was estimated by the modified Egger linear regression test. A 95% CI not crossing 1 was considered significant. All analyses were performed using the software R-3.3.1 .The Chi-squared test was used to assess the deviation of genotype distribution from HWE among controls. Crude ORs and corresponding 95% CIs were computed to assess the relationship between IL-4 and IL-4R polymorphisms and periodontitis susceptibility. The pooled ORs were calculated for the allele contrast, codominant, dominant, and recessive model, respectively. Heterogeneity was quantified using the As shown in Due to significant heterogeneity detected Tables , and 4, https://doi.org/10.1155/2017/8021279.Meta-analysis of the IL-4 -590C/T showed no association between the polymorphism and periodontitis susceptibility . Subgroup analyses by disease type, ethnicity, and HWE status for controls were similar to the overall analyses . AccordiI2 = 0%) . Subgroup analyses according to disease type showed negative association with AgP  with no between-study heterogeneity (I2 = 0%) . AccordiMeta-analysis of the IL-4 -1099T/G showed no association between the polymorphism and periodontitis susceptibility  . SensitiDue to significant heterogeneity detected , the ranI2 of the 12 versus 11 model decreased to 0% after its removal. The pooled result was insensitive to other individual studies. The details of sensitivity analysis were shown in supplementary table 1.Meta-analysis of the IL-4 70-bp VNTR showed no association between the polymorphism and periodontitis susceptibility . The results of stratification analyses according to disease type and ethnicity were similar to the overall results . SensitiNo significant heterogeneity was detected ; thus fiMeta-analysis of the IL-4R Q551R polymorphism showed a positive association between the polymorphism and periodontitis susceptibility in three genetic models  with low between-study heterogeneity . SubgrouP values were P = 0.01 and P = 0.02, respectively, for T versus C allele model and TT versus (CC + CT) model. There is no publication bias in the other three genetic models.Due to limitations of the quantity of included studies, we only tested the publication bias for IL-4 -590C/T polymorphisms. The funnel plots for T versus C allele model and TT versus (CC + CT) model suggested that there probably existed publication bias. Egger's test showed In the present study, we performed meta-analyses concerning the relationship between susceptibility and periodontitis (including CP and AgP) and polymorphisms including the IL-4 -590C/T, -33C/T, -1099T/G, 70-bp VNTR, and IL-4R Q551R, in the overall population and specific subgroups. Our study showed that the IL-4R Q551R R allele could significantly increase susceptibility to periodontitis in Caucasians, especially in terms of CP, which was more evident in those having a history of tobacco smoking. Similar findings were revealed with respect to the QR versus QQ and RR + QR versus QQ comparisons. The IL-4 -33C/T polymorphism was not associated with periodontitis susceptibility in the overall population; however, stratified analysis by type of disease showed that the CT genotypes were negatively associated with AgP. However, this association was nonsignificant as to CP. According to the results of overall meta-analyses and subgroup analyses by ethnicity, type of disease, HWE, and smoking status, no significant association was found between periodontitis susceptibility and polymorphisms IL-4 -590C/T, -1099T/G, and 70-bp VNTR, with the only exception that, in terms of the IL-4 -590C/T polymorphism, Caucasians with TT genotype showed marginally significant trend toward having periodontitis .To the best of our knowledge, for the first time the IL-4 -1099T/G and IL-4R Q551R polymorphisms were included for meta-analytically analyzing the association between periodontitis susceptibility and gene polymorphisms, and the present study is the most comprehensive synthesis concerning polymorphisms on IL-4 and IL-4R and susceptibility to periodontitis. According to our results, it is demonstrated that variety exists in terms of IL-4 and IL-4R polymorphisms' distribution among different regions, populations, and disease types, and new evidence is produced. Compared to previously published meta-analysis, there are more studies included in the present meta-analysis, and the overall sample size is larger; therefore, our findings are more reliable. In addition, two more polymorphisms, that is, the IL-4-1099T/G and IL-4R Q551R, are discussed in our study, and it is revealed that the IL-4R Q551R allele may increase the susceptibility to periodontitis in Caucasians. Considering the complexity of the etiology and gene distribution of this disease, we performed comprehensive subgroup analysis stratified by ethnicity, disease type, and smoking status, by which our study topic was thoroughly analyzed.Nevertheless, our study has some limitations. Firstly, although comprehensive literature search was conducted, certain gray literature could not be obtained, and therefore potential publication bias could not be excluded. Secondly, 5 studies inconsistent with HWE were included in our study, though sensitivity analysis found no significant influence on the overall results, and potential population bias could not be excluded. Significantly, major heterogeneity was found in most of the meta-analyses. Reduced heterogeneity was observed in some of the subgroup analyses. Given the complexity of periodontitis and potential confounding factors such as psychological stress and number of lost teeth, difference in clinical and/or environmental factors might have contributed to the heterogeneity among individual studies. Finally, the number of available studies was very limited in some subgroup analyses, and, due to the limited sample size, the pooled results were less accurate and more studies with large sample size and high quality are needed for further investigation.To summarize, the present meta-analysis has demonstrated that a positive association exists between the IL-4R Q551R polymorphism and occurrence of CP. The IL-4 -33 CT genotype is negatively associated with the occurrence of AgP. However, this is not the case for other polymorphisms discussed. Our study findings provide guidance for early diagnosis and treatment of this disease. However, due to the limited number of studies included, additional studies with large sample size and adequate quality are needed to further verify and confirm our findings. Besides, interactions between multiple polymorphisms and between genetic and environmental factors should be investigated.In Supplementary Table 1 are shown collected resuts of sensitivity analyses. Individual influential studies are identified.


Objective: Several studies have shown that some polymorphisms of genes encoding catechol-O-methyltransferase (COMT), the key enzyme in degrading dopamine, and norepinephrine and the human brain-derived neurotropic factor (BDNF), a nerve growth factor, are strong candidates for risk of schizophrenia (SCZ). In the present study, we aimed at examining the effects of COMT Val158Met (G>A) and BDNF Val66Met (G>A) polymorphisms on SCZ risk in a sample of Iranian population.Method: This case- control study included 92 SCZ patients and 92 healthy controls (HCs). Genotyping of both variants  and BDNF Val66Met (G>A)) were conducted using Amplification Refractory Mutation System-Polymerase Chain Reaction (ARMS-PCR).Results: The findings revealed that the COMT Val158Met (G>A) polymorphism was not associated with the risk/protective of SCZ in all models . However, BDNF Val66Met (G>A) variant increased the risk of SCZ .Conclusion: The results did not support an association between COMT Val158Met (G>A) variant and risk/protective of SCZ. Moreover, it was found that BDNF Val66Met (G>A) polymorphism may increase the risk of SCZ development. Further studies and different ethnicities are recommended to confirm the findings. Schizophrenia (SCZ) is a complex psychiatric disorder that affects nearly 1% of the general population worldwide characterized by hallucinations, delusions, and incompatibility of mental activity and environment; however, the exact mechanisms underlying the development and progression of SCZ are still entirely unknown, but genetic factor accounts for more than 80% of etiology and heritability of SCZ .The human Catechol-O-methyltransferase (COMT) gene is located on the long arm of Chromosome 22 at Position 11.21 (22q11) microdeletion, COMT with transferring a methyl group to catecholamine due to their degrading including dopamine, adrenaline, and noradrenaline , 5. cted on the association of COMT gene as a candidate gene for SCZ, so the activity of this enzyme is regulated by a common polymorphism causing substantial variations in the activity of this enzyme , 6, 7.A functional single nucleotide polymorphism (SNP) with code rs4680G>A causes guanine change to adenine (exchange between nucleotide) in the gene sequence, which produces methionine amino acid from valine amino acid  at Codon 158. This change, which is due to the reduction of the activity of COMT enzyme with Met vs. Val (one-fourth of the enzymatic activity), results in a slower degradation of dopamine and greater availability , 10. MorThis case-control study was performed on 92 patients with SCZ who referred to Baharan hospital in the city of Zahedan, Iran from June 2015 to June 2016). Their disease was diagnosed by well-trained psychiatrists according to the Diagnostic and Statistical Manual of Mental Disorder, fourth edition  and previous works of the authors , 19. A tEthicsThe local ethics committee of Zahedan University of Medical Sciences  approved the study protocol. Informed consent was obtained from all participants (and the next of kin) before participating in this study.Genomic DNA ExtractionGenomic DNA was extracted from whole blood (2mL in EDTA-containing tubes) of the patient and control groups immediately after collection using salting-out method as described previously . The elePCR ReactionThe analysis of COMT Val158Met (G>A) and BDNF Val66Met (G>A) polymorphisms were performed by Amplification Refractory Mutation System- Polymerase Chain Reaction (ARMS-PCR). The list of primers, designed by using Wasp primer software,  used for amplification of SNPs is provided in Gel ElectrophoresisPCR products were verified on 1.5% of agarose  containing ethidium bromide (0.7μg/ml)  observed under UV light. The sizes of PCR products were reported to be 244bp for COMT Val158Met (G>A) and 258bp for BDNF Val66Met (G>A). At least, 20% of samples, either SCZ or HCs, were regenotyped randomly, and the new results confirmed the past results.Statistical AnalysisX2 test. A probability of p<0.05 was considered statistically significant.Statistical analyses were performed using SPSS statistical software Version 22 . The rates of the genotypes and alleles were compared using the binary logistic regression and Mann-Whitney test. Categorical data were compared using Pearson’s The demographic characteristics of the patients with SCZ and HCs are provided in The frequencies of alleles and genotype of COMT Val158Met (G>A) and BDNF Val66Met (G>A) polymorphisms are presented in In addition, BDNF Val66Met (G>A) in the dominant (GA+AA vs. GG) model indicated an increased risk of SCZ (OR = 2.272. 95%CI = 1.204-4.347. P = 0.011). As demonstrated in X2 = 3.27, P = 0.07 and X2 = 1.74, P = 0.19, respectively), another SNP, COMT Val158Met (G>A), did not confirm HWE .The BDNF Val66Met (G>A) polymorphism conformed to the Hardy Weinberg Equilibrium (HWE) in the SCZ patients and HCs  and SCZ, while the BDNF Val66Met (G>A) variant was associated with the risk of SCZ, either genotype or allele. SCZ is a multifactorial disease with unclear and unknown etiology, but several factors are possibly involved in the SCZ incidence, prevalence, and mortality such as sex, migration status, urban status, economic status, and latitude . It is sCOMT enzyme is an important degradation enzyme involved in the metabolism of catecholamine, consisting of dopamine, adrenaline, and noradrenaline. Evidence shows that polymorphism of COMT gene may influence the activity of COMT enzyme; consequently, the effectiveness of COMT enzyme decreases, and slower degradation of dopamine may be involved in SCZ development , 22. AnoSimilarly, several investigations have indicated an association between BDNF Val66Met (G>A) polymorphism and SCZ, confirming the results of the current study. Naoe et al. performed a meta-analysis on 8 case control studies including 2156 Caucasians with SCZ and 3007 HCs and suggested that Val66Met (G>A) polymorphism of BDNF gene was significantly associated with SCZ risk . Neves-Pa case-control study performed by Chen et al. on Han Chinese people, BDNF Val66Met (G>A) polymorphism was not associated with SCZ development . Also, WContradictory to the current results, Tochigi et al. did not find any association between BDNF Val66Met  (G>A) variant and risk of SCZ susceptibility in Japanese population . Chen-jeVal66Met (G>A) and SCZ in several case-control and meta-analysis studies in Asia , 26, 27.Several studies have also shown no association between COMT Val158Met (G>A) polymorphism and SCZ, approving the findings of the present study. Daniels did not find any correlation between COMT Val158Met (G>A) and SCZ in Wales . SimilarIn our population, Hardy-Weinberg disequilibrium could be due to several reasons such as the structure of population in our area (different ethnic groups), migration, small sample size, and consanguineous marriages.The current investigation had several limitations. First, based on the published and Medline data, several COMT and BDNF polymorphisms have been identified in humans, while we investigated only 1 polymorphism from each gene. Second, we did not consider the differences according to specific schizophrenia subtypes. Third the small sample size.In conclusion, our results revealed a significant association between BDNF Val66Met (G>A) polymorphism and SCZ risk, not supporting an association between COMT Val158Met (G>A) and risk of SCZ in the population of Southeast Iran. Further studies with larger sample sizes and different ethnicities should be conducted to determine their effects on SCZ risk.


Previous studies have revealed the importance of microRNAs’ (miRNAs) function as biomarkers in diagnosing human bladder cancer (BC). However, the results are discordant. Consequently, the possibility of miRNAs to be BC biomarkers was summarized in this meta-analysis.In this study, the relevant articles were systematically searched from CBM, PubMed, EMBASE, and Chinese National Knowledge Infrastructure (CNKI). The bivariate model was used to calculate the pooled diagnostic parameters and summary receiver operator characteristic (SROC) curve in this meta-analysis, thereby estimating the whole predictive performance. STATA software was used during the whole analysis.Thirty-one studies from 10 articles, including 1556 cases and 1347 controls, were explored in this meta-analysis. In short, the pooled sensitivity, area under the SROC curve, specificity, positive likelihood ratio, diagnostic odds ratio, and negative likelihood ratio were 0.72 (95%CI 0.66–0.76), 0.80 (0.77–0.84), 0.76 (0.71–0.81), 3.0 (2.4–3.8), 8 (5.0–12.0), and 0.37 (0.30–0.46) respectively. Additionally, sub-group and meta-regression analyses revealed that there were significant differences between ethnicity, miRNA profiling, and specimen sub-groups. These results suggested that Asian population-based studies, multiple-miRNA profiling, and blood-based assays might yield a higher diagnostic accuracy than their counterparts.This meta-analysis demonstrated that miRNAs, particularly multiple miRNAs in the blood, might be novel, useful biomarkers with relatively high sensitivity and specificity and can be used for the diagnosis of BC. However, further prospective studies with more samples should be performed for further validation. Bladder cancer (BC) as the second common urinary system malignancy is a leading cause of global cancer-related deaths. The incidences of BC were 37.5 per 100,000 in men and 9.3 per 100,000 in women , 5. In aAt present, there are a number of conventional diagnostic methods for BC detection. These include cystoscopy, voided urine cytology, and detection of several urine-based markers such as NMP22 (nuclear matrix protein 22) and BTA (bladder tumor antigen). The most common way is cystoscopy coupled with voided urine cytology, which is highly sensitive , 9. HowemiRNAs are a series of short , single-stranded, and non-protein-coding RNA gene products that are vital for the regulation of gene expression. They work through binding their target mRNAs to cause degradation or translational silencing . Recent The relevant literature in the ExcerptaMedica Database (EMBASE), PubMed, Chinese Biomedical Literature (CBM) database, and Chinese National Knowledge Infrastructure (CNKI) web database  were searched using a combination of the terms “microRNAs” or “miRNA” or “miR”, “bladder cancer” or “bladder tumor” or “bladder urothelial cell carcinoma” and “sensitivity” or “specificity”, “ROC curve” or “diagnosis”. In order to retrieve the most relevant studies, manual filtration of the references was also conducted.The adopted literature should meet the following criteria: (a) the diagnostic value of miRNAs for BC detection must have been evaluated, (b) the authors must have used the gold standards , and (c) the studies must contain detailed information for constructing two-by-two tables, which comprise true positives (TP), true negatives (TN), false positives (FP), and false negatives (FN). And exclusion criteria are as follows: (a) abstract, review, comment, editorial, and case reports, (b) overlapping data, (c) studies investigating survival or prognosis of BC, (d) deficient data, even on asking the corresponding authors for more information, and (e) sample size <100. If more than one study used some common samples, only the most precise one with the most samples was selected.Data extraction from the selected publications was done using a standardized table by our two authors independently. And the extracted data contains first author’s surname, publication time, and country of origin, sample size, age, type of case, and control groups, sample specimen, studied miRNAs, detection method, TP, FP, FN, and TN, and information needed for quality assessment. For studies involving more than one type of miRNAs or sample specimen, the data were extracted, and the study on each miRNA was considered as an independent study. The quality of each article was evaluated by the revised quality assessment of diagnostic accuracy studies (QUADAS-2) checklists . All disQ and I2 tests. A P value <0.10 and an I2 value >50% were considered as significant heterogeneity [P value <0.05 [P values were two sided and all statistical analyses in this meta-analysis were performed by using the STATA statistical software .For purpose of assessing the diagnostic value of miRNAs for BC, a bivariate meta-analysis was managed to obtain pooled negative likelihood ratio (NLR), positive likelihood ratio (PLR), sensitivity, diagnostic odds ratio (DOR), specificity, and corresponding 95% confidence intervals (CIs) . Meanwhiogeneity , 28. In ue <0.05 . All P vThere were 370 relevant articles picked out after preliminary screening in databases. The flow chart illustrates the study filtration procedures used for assessing the diagnostic potential of miRNAs in BC Fig. . After aThe summarized estimates of the diagnostic accuracy of miRNAs that distinguish BC from controls are listed in Table Sub-group analyses and meta-regression were conducted in order to reveal the inter-study heterogeneity attributed to ethnicity, miRNA profiling, specimens, and sample size Table . The resP value <0.01), ethnicity , miRNA profiling , and specimen  could be the reasons of the heterogeneity in sensitivity and specificity between the studies  to the bivariate model. The meta-regression analysis suggested that sample size . All results confirmed that there was no obvious publication bias.Deeks’ funnel plot asymmetry test was used to evaluate the possible publication bias in the included studies. Overall, the funnel plots showed symmetry and the slope coefficient was associated with a .11 Fig. . For theThis meta-analysis is the first evidence-based analysis to assess the potential role of miRNAs for BC detection. This meta-analysis includes 31 studies from 10 researches covering 1556 cancer cases and 1347 controls in all, and the results indicate that miRNAs are potential novel and useful BC biomarkers with relatively high sensitivity and specificity that can be used for BC diagnosis.Due to the high morbidity in men, high mortality for MIBC, and frequent recurrence, aggressive surveillance and timely intervention can dramatically improve the prognosis of BC. Currently, the method of the initial clinical diagnosis of BC—cystoscopy coupled with voided urine cytology—, 9 is exIt should be noted that available heterogeneity existed between these studies used in this paper; therefore, the effects of these confounding factors were explored via sub-group analyses. When stratified by miRNA profiling, the interesting finding was that single-miRNA assays showed relatively poor diagnostic performance with an AUC value of 0.74, while the diagnostic accuracy of multiple-miRNA assays is significantly higher with an AUC of 0.92. The results indicate that combination of multiple miRNAs might be more useful for the diagnosis of BC than single miRNAs. In addition, numerous studies have validated that single miRNAs can be used as accurate biomarkers for cancers. For example, Yang et al. performed a meta-analysis to evaluate the diagnostic performance of miRNA-21 in serum for lung carcinoma detection and suggested that miRNA-21 has diagnostic potential with an AUC value of 0.86 . AnotherWhen stratified by sample types, the results indicate that comparing to urine-based assays, blood-based assays could be a better choice as a diagnostic tool. Ding et al. carried out a meta-analysis to evaluate the diagnostic value of miRNAs for pancreatic cancer and made a similar observation that blood-based assays were significantly better than non-blood-based assays . HoweverFurthermore, the sub-group analysis by ethnicity suggests that the diagnostic accuracy of miRNA assays is significantly better in Asian populations than that in Caucasian populations, with a pooled DOR of 25.0 vs. 6.0 and AUC of 0.90 vs. 0.76, respectively. Li et al. conducted a meta-analysis to evaluate the diagnostic value of miRNAs for hematologic malignancies and found that Asian population-based miRNA tests yield an overall higher accuracy than tests in Caucasian populations . HoweverThis study has several strengths notwithstanding a few limitations. This is the first meta-analysis to use as many as 1556 cancer cases and 1347 controls to summarize the diagnostic value of miRNAs in BC, which gives improved statistical power to the findings. The sub-group analysis confirmed that Asian population-based studies, multiple-miRNA profiling, and blood-based assays might yield increased diagnostic accuracy. Finally, this meta-regression analysis aimed to explore the sources of heterogeneity indicated that ethnicity, miRNA profiling, and sample type were independent confounding factors. However, the limitations of this study should also be acknowledged. Firstly, the cut-off parameters for miRNAs were different in the studies used and could be a potential source of heterogeneity. Secondly, this meta-analysis did not evaluate differences in the diagnostic accuracy of miRNAs presented in BCs with different clinicopathological features.This meta-analysis provides the first evidence of miRNAs being novel, useful biomarkers for the diagnosis of BC. Detection of miRNAs can be done with high sensitivity and specificity, particularly using multiple-miRNA blood-based assays. However, extensive functional evaluations and further population-based prospective studies with larger sample sizes and different ethnic groups are warranted to confirm and extend these findings.


P = 0.019; P < 0.001 for heterogeneity, I2 = 72.7%). When stratified by study design, the RRs for case-control and cohort studies were 0.76  and 0.96 , respectively. In the subgroup analysis by geographical region, significant protective effects were observed in Asian  and Oceania populations , but not in other geographical populations. Begg’s test indicated a significant publication bias (P = 0.015). Overall, tomato intake may have a weak protective effect against prostate cancer. Because of the huge heterogeneity and null results in cohort studies, further prospective studies are needed to explore the potential relationship between tomato consumption and prostate cancer risk.Previous studies have reported controversial results on the association between tomato consumption and prostate cancer risk. Hence, we performed a meta-analysis to comprehensively evaluate this relationship. A total of 24 published studies with 15,099 cases were included. Relative risks (RR) and 95% confidence intervals (CI) were pooled with a random-effects model. Tomato intake was associated with a reduced risk of prostate cancer (RR 0.86, 95% CI 0.75–0.98,  Emerging evidence from epidemiological, as well as cell culture and animal, studies indicates that lycopene and the consumption of lycopene-containing foods may be protective against cancer and cardiovascular disease risk4et al.Processed tomato products are the primary dietary lycopene source789101112131416171819A previous meta-analysis published in 2004 reported that tomato consumption might play a protective role in the prevention of prostate cancer based on three cohort and seven case-control studies7891011121314151617181920212425262728293031P = 0.019). Statistically significant heterogeneity was observed among included studies .Multivariable adjusted relative risks (RRs) with their confidence intervals (CIs) for each individual study and for the combination of all included studies are shown in P = 0.015) and Oceania populations , but the effects were not significant in other geographical populations. When stratified by study design, the analysis of case-control studies yielded a RR of 0.76 , whereas the analysis based on cohort studies yielded a RR of 0.96  (P = 0.030) compared with high-quality studies . Finally, in the stratified analyses by sample size, statistically significant association was observed in those small studies  rather than in large studies .The effects of tomato intake on prostate cancer risk in subgroup meta-analyses are shown in = 0.579) . In the et al.et al.P = 0.015), but not in Egger’s test (P = 0.122).The influence of each study on the pooled RR was evaluated by repeating the overall analysis after omitting each study in turn. The results indicated that no single study dominated the combined RR. The 24 study-specific RRs ranged from a low of 0.83 (95% CI 0.72–0.97) to a high of 0.89 (95% CI 0.79–1.00) via omission of the study by Stram This systematic review and meta-analysis aimed to evaluate the relationship between tomato intake and prostate cancer risk based on 7 cohort studies and 17 case-control studies, with a total of 15,099 cases. The results of this quantitative meta-analysis provided limited evidence for a protective effect of high tomato food consumption for prostate cancer incidence. Although the overall analysis suggested a moderate reduction in risk, the results from the cohort, high-quality, and large studies were null.I2 = 52.0%)I2 = 54.1). Therefore, a protective effect of tomato intake on the risk of prostate cancer is mainly observed in case-control studies. Compared with these previous meta-analyses, the present updated meta-analysis also performed a stratified analysis by geographical region, which provided a more comprehensive assessment of the association between tomato consumption and prostate cancer risk.The findings of this meta-analysis are basically consistent with a previous meta-analysis published in 2004et al.et al.In vivo studies, dietary tomato and lycopene could have an influence on androgen signaling- and carcinogenesis-related gene expression during early transgenic adenocarcinoma of the mouse prostate (TRAMP) mice prostate carcinogenesisSeveral potential mechanisms could explain the potential cancer-protective effects of tomato food. Tomato food has high levels of lycopene, which has been shown to inhibit prostate cancer progression in several studies. Yang This study had several important strengths. First, as individual studies may have limited statistical power, our meta-analysis of 24 published studies with 15,099 prostate cancer cases might provide more reliable results with greater precision and power. Second, we extracted data from the most fully adjusted model in each study, which reduce the potential influence of confounding factors. Third, various subgroup analyses, influence analysis, and publication bias analysis were performed to evaluate the robustness of the pooled risk estimate.P < 0.001 for heterogeneity, I2 = 72.7%), which was likely due to the variation in population information, exposure definitions, exposure ranges, exposure and outcome assessment methods between studies. Second, Begg’s test suggested the existence of publication bias. Although we adopted a loose search strategy, some inevitable publication bias might exist as small studies with negative results were less likely to be published and gray literature (such as non-English articles and conference abstract) was difficult to find. Third, the cutoff points for the lowest and highest categories of the tomato intake were various in included studies, which might also has an influence on the combined risk estimate. Finally, the association between lifestyle factors and prostate cancer risk may vary by tumor characteristics . However, most of the included studies didn’t provide risk estimates for localized/low grade and advanced/high grade cancers separately. Therefore, we were not able to examine if there were differences by stage and grade in the association between tomato intake and prostate cancer risk.However, several limitations should be considered in interpreting the results of this meta-analysis. First, there was substantial heterogeneity across studies  and (“prostatic neoplasms” or “prostatic cancer” or “prostate neoplasms” or “prostate cancer”). Furthermore, the cited references of retrieved articles and reviews were also checked to identify any additional relevant studies. There was no language, publication date, or publication status restrictions. This systematic review and meta-analysis was designed, performed, and reported in accordance with the standards of quality for reporting meta-analyses, except for not publishing the review protocol in advancei) the exposure of interest was consumption of tomato food; (ii) the outcome of interest was incidence of prostate cancer; (iii) study design was cohort, nested case-control or case-control; and (iv) the effect sizes with their corresponding 95% CIs were reported. If multiple articles reported data based on the same population, the publication with the most up-to-date or comprehensive information was included in the meta-analysis.A study was included if it met the following criteria:  was used to assess the quality of each included study by two independent reviewers (XX and JFL). NOS judges a study according to the following three broad perspectives: selection (four items), comparability (one item), and exposure/outcome (three items). Each item is awarded one point, except for comparability (two points). Hence, the full score is 9 stars. A study with ≥7 awarded stars is classified as high quality.A 9-star system on the basis of the NOS (Information was collected and recorded independently by two investigators (XX and JFL). Any discrepancies were resolved through iteration and consensus. The following data were obtained from each study: first author’s surname, country, publication year, study design, age, number of cases, instrument of exposure measurement, method of outcome assessment, results of studies (adjusted risk estimates with their corresponding 95% CIs), and matched or adjusted confounding factors in the design or statistical analysis.Considering that prostate cancer is a rare disease, the odds ratio (OR) was assumed approximately the same as RR, and the RR was designated as the study outcome. Multiple adjusted RRs with their 95% CIs were used to measure the strength of the relationship between tomato intake and prostate cancer risk. Some studies reported risk estimates for raw tomato and cooked tomato separately and did not report the effect of total tomato intake. In this situation, the study-specific RR in overall analysis was recalculated by pooling the risk estimates with the inverse-variance methodQ test and the I2 scoreQ was test set at 0.1 (10%). The I2 score was adopted to evaluate the degree of heterogeneity .Statistical heterogeneity among included studies was estimated using Cochran’s P values (set at 0.05).A sensitivity analysis was conducted by omitting each study in turn and recalculating the pooled RR to test the impact of each study on the overall risk estimate. Potential publication bias was assessed through Begg’s test (rank correlation method)How to cite this article: Xu, X. et al. Tomato consumption and prostate cancer risk: a systematic review and meta-analysis. Sci. Rep.6, 37091; doi: 10.1038/srep37091 (2016).Publisher’s note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


Observational studies examining associations between adult height and risk of colorectal, prostate, and lung cancers have generated mixed results. We conducted meta-analyses using data from prospective cohort studies and further carried out Mendelian randomization analyses, using height-associated genetic variants identified in a genome-wide association study (GWAS), to evaluate the association of adult height with these cancers.A systematic review of prospective studies was conducted using the PubMed, Embase, and Web of Science databases. Using meta-analyses, results obtained from 62 studies were summarized for the association of a 10-cm increase in height with cancer risk. Mendelian randomization analyses were conducted using summary statistics obtained for 423 genetic variants identified from a recent GWAS of adult height and from a cancer genetics consortium study of multiple cancers that included 47,800 cases and 81,353 controls. For a 10-cm increase in height, the summary relative risks derived from the meta-analyses of prospective studies were 1.12 , 1.07 , and 1.06  for colorectal, prostate, and lung cancers, respectively. Mendelian randomization analyses showed increased risks of colorectal  and lung cancer  associated with each 10-cm increase in genetically predicted height. No association was observed for prostate cancer . Our meta-analysis was limited to published studies. The sample size for the Mendelian randomization analysis of colorectal cancer was relatively small, thus affecting the precision of the point estimate.Our study provides evidence for a potential causal association of adult height with the risk of colorectal and lung cancers and suggests that certain genetic factors and biological pathways affecting adult height may also affect the risk of these cancers. In a Mendelian randomisation study Pierce and colleagues show a genetic association between adult height and increased risk of colorectal and lung cancer. Several previous observational studies have examined the association between adult height and risk of cancers of the lung, colon/rectum, and prostate; however, it remains unclear whether adult height is indeed related to the risk of these cancers.We conducted a systematic review and meta-analysis of prospective cohort studies that examined the association between adult height and the risk of colorectal, lung, and prostate cancers.To overcome inherent limitations of observational study designs, we conducted Mendelian randomization analyses using genetic data generated from a large multi-center consortium study including 47,800 cases and 81,353 controls.In the meta-analysis of the prospective observational studies, we found a 12% increased risk of colorectal cancer, a 7% increased risk of prostate cancer, and a 6% increased risk of lung cancer for every ten-centimeter increase in height, and this increased risk was corroborated in the Mendelian randomization analyses for colorectal (58%) and lung cancer (10%).Our study provides strong evidence for an association between adult height and risk of colorectal and lung cancer, and suggests that certain genetic and biological factors that affect height may also affect the risk of these cancers.However, our meta-analysis was limited to published studies, and the sample size for the Mendelian randomization analysis for colorectal cancer was relatively small, affecting the precision of the risk estimate. Numerous studies have examined the relation between height and cancer; however, results have been inconsistent. The inconsistencies in observational studies could be due to factors that affect validity, including confounding, selection bias, reverse causation, and measurement error. Therefore, no consensus has been reached on whether height is a risk factor for colorectal, lung, or prostate cancer, three of the most common cancers affecting men and women. Furthermore, meta-analyses of prospective cohort studies, summarizing the association between height and colorectal, lung, or prostate cancers, have not been conducted.Mendelian randomization is based on the principle that an individual’s genotype is randomized at conception  and utilTo comprehensively evaluate the association between height and risk of colorectal, lung, and prostate cancers, we conducted a systematic review and meta-analysis of previous prospective studies. Additionally, we carried out Mendelian randomization analyses utilizing GWAS summary statistics from the Genetic Associations and Mechanisms in Oncology (GAME-ON) and Genetic Investigation of Anthropometric Traits (GIANT) consortia studies of individuals of European descent.We searched PubMed, Embase, and Web of Science on January 3, 2016, for prospective studies  using the following search terms: lung neoplasms; lung cancer; colorectal neoplasms; colorectal cancer; prostate neoplasms; prostate cancer; lung, prostate, colorectal, colon, rectum, rectal or cancer; and body height, height, stature, body size, anthropometrics, or anthropometry. After restricting the results to English language and humans, a total of 15,691 publications were found.Approximately 2,900 publications were duplicates. Following the title and abstract review, additional exclusions for outcomes other than cancer resulted in a total of 325 studies for full-text review. One Chinese study, for which all relevant tables and methods were written in English, was additionally included . Full-teAll study estimates were scaled for a 10-cm increase in height in relation to cancer, which, in general, approximated the average interquartile range in height across all studies. For studies that reported categorical estimates of height , a continuous estimate and standard error was estimated using the method proposed by Greenland and Longnecker  using R p < 0.05 indicating heterogeneity among the studies. Summary odds ratios (ORs) and 95% CIs for the random-effects meta-analyses were estimated using restricted maximum likelihood. Funnel plots were generated for each cancer to examine publication bias and were formally assessed using Begg’s and Egger’s statistical tests (Study-specific estimates were summarized using inverse-variance-weighted (IVW) fixed- and random-effects meta-analyses for all three cancers  separately. Fixed-effects models assume no heterogeneity between the study estimates, whereas random-effects models take heterogeneity of the study estimates into consideration. Cochran’s test for homogeneity was applied in the analysis, with al tests –S4 Figs;X) on the risk of cancer (Y) using genetic variants (g), where the causal estimate is equal to Yg/Xg [Xg), we utilized available summary statistics from the GIANT GWAS. Summary statistics for the association between genetic variants and each of the three cancers (Yg) are from the GAME-ON consortium.Results from our recent Mendelian randomization analysis of adult height and breast cancer have been published , and thuYg/Xg . For thep < 5 × 10−8) in the GIANT consortium study, explaining 16% of phenotypic variation in height [Nearly 700 genetic variants from 423 loci were identified as reaching genome-wide significance p <  × 10−8 iYg). GAME-ON is network of genetic epidemiology consortia that contains GWAS data from the Colorectal Transdisciplinary Study (CORECT); Discovery, Biology, and Risk of Inherited Variants in Breast Cancer (DRIVE); Elucidating Loci Involved in Prostate Cancer Susceptibility (ELLIPSE); Follow-up of Ovarian Cancer Genetic Association and Interaction Studies (FOCI); and Transdisciplinary Research in Cancer of the Lung (TRICL). For the current study, we used data from CORECT, ELLIPSE, and TRICL. The GAME-ON consortium primarily consists of studies of individuals of European descent. SNPs with poor imputation quality (r2 < 0.3) were excluded (CORECT used IMPUTE 2.0 info score < 0.7 as the cutoff) [GAME-ON data were utilized to provide regression coefficients and standard errors for each single nucleotide polymorphism (SNP) in relation to cancer ( cutoff) . The ass cutoff) . For theXg (from the GIANT consortium) had corresponding information regarding Yg (from GAME-ON); therefore, these SNPs were excluded from the instrumental variables for colorectal cancer (n = 77) and prostate cancer (n = 4). No SNPs were excluded from the instrument for lung cancer. Thus, there were fewer instrumental variables for the colorectal and prostate cancer studies than for the lung cancer study. Using both summary statistics for Yg and Xg, an IVW meta-analysis was conducted to estimate the effect of genetically determined height on the risk of each cancer using the method of Burgess et al. [Xg is the beta estimate for the association between the SNP and height (from GIANT), Yg is the beta estimate for the association between the SNP and cancer (from GAME-ON), and σYg is the standard error for Yg. Corresponding ORs and 95% CIs were calculated using Not all SNPs with information regarding s et al. :β^IVW=∑We conducted several sensitivity analyses to assess whether the association of height with cancer varied for different subgroups. Using data from a previous GIANT height GWAS that presented sex-stratified summary statistics for 168 independent SNPs explaining approximately 10% of variation in height , we condp < 0.001); no important differences were observed when stratified by sex. A 7% increase  in the risk of prostate cancer  and those that did not . Similarly, no differences were observed for smoking-adjusted studies for colorectal cancer . For prostate cancer, a modest difference was observed , but the difference was not statistically significant. Studies using self-reported height and studies using measured height yielded very similar summary estimates for prostate cancer  and colorectal cancer . However, some modest differences were observed for lung cancer . Additionally, the random-effects summary estimates for colorectal, prostate, and lung cancers did not differ substantially when we restricted the studies to only those conducted among populations of European descent  than that for colon cancer . No apparent differences were observed for lung cancer between studies that adjusted for smoking (RR descent .p = 0.006); males and females had slightly different estimates per 10-cm increase in height, with females having a slightly higher risk than males (p = 0.053). The association appears to be stronger for adenocarcinoma of the lung  than for squamous cell carcinoma of the lung , and the homogeneity p-value was not statistically significant (p = 0.427). However, the estimates for males and females according to subtype were nearly identical . The results for breast cancer showed an approximately 20% increase in risk per 10-cm increase in genetically predicted height , which is very similar to our previous estimate using a smaller set of height-associated genetic variants  [Results from the Mendelian randomization analyses are presented in an males . A modesdentical . No stats p = 0.0. The assWe did not observe any substantial differences in the Mendelian randomization estimates after excluding genetic variants with suggested pleiotropic effects from the instrument . FurtherIn this meta-analysis involving 62 published prospective studies, we observed a statistically significant increased risk for colorectal, prostate, and lung cancers associated with higher adult height. Our Mendelian randomization analyses confirmed these increased risks for colorectal and lung cancers but not for prostate cancer, indicating perhaps that the increased risk reported from previous prospective cohort studies for prostate cancer may be due to biases. To our knowledge, this is the first meta-analysis performed to summarize results from prospective cohort studies regarding the association of adult height and these three major cancers that affect the lives of men and women throughout the world. To our knowledge, this is also the first Mendelian randomization analysis performed to evaluate the association between adult height and risk of lung cancer. Our study, using an instrument comprising 423 height-related SNPs and explaining approximately 16% of phenotypic variation, provides strong evidence for a possible causal association between adult height and risk of breast, colorectal, and lung cancers. It suggests that certain genetic factors and biological pathways affecting adult height may affect the risk of these major cancers.One potential reason taller individuals may be at higher risk is the increased opportunity for a cell to mutate given that taller people have more cells. However, multiple interrelated biological pathways have also been implicated in the association between height and cancer. One pathway is the insulin-like growth factor (IGF) pathway, which is known to promote cell proliferation and inhibit apoptosis ; severalSMAD3, MTOR, GLI2, LAMA5) involved in these pathways were related to height in a recent GWAS [It is unlikely that the IGF-1 pathway alone would explain entirely the observed increased colorectal and lung cancer risk associated with adult height. Other pathways have also been revealed recently to influence adult height, including transforming growth factor beta (TGF-β)  and HedgThe results are similar when comparing the summary estimates from the meta-analysis and the associations from the Mendelian randomization analysis for both breast and lung cancers. For colorectal cancer, however, the association from our Mendelian randomization analysis was substantially stronger than that estimated from the meta-analysis of prospective cohort studies. This difference is likely due to sampling errors, given the small sample size of our Mendelian randomization analysis performed for colorectal cancer. Indeed, in a recent Mendelian randomization study with a larger sample size  than our study, a 10-cm increase in height was associated with a 7%  increased risk of colorectal cancer, and the risk was higher among females  than males  . AlthougWe report modest increases in the risk of lung cancer with height in both the meta-analysis of prospective studies and the Mendelian randomization analysis. The majority of lung cancers are caused by smoking, and smoking may reduce IGF-1 levels . Future r2 ≥ 0.3 for ELLIPSE and TRICL, and info score ≥ 0.7 for CORECT) [n = 36) with suggested pleiotropic effects on several different biological outcomes, including pulmonary function, sex hormone binding globulin levels, and age at menarche. Pleiotropic effects would violate the exclusion restriction assumption for a valid instrument, which states that the instrument affects the outcome only through the exposure phenotype of interest [There are several strengths to this study. The comprehensive literature search identified 62 studies related to adult height and the cancers evaluated in our study. Additionally, comparing the results from the Mendelian randomization analysis with the meta-analysis summary estimates could indicate whether or not the results from observational studies are biased, assuming the Mendelian randomization analyses represent an effect that may be closer to the true effect. Furthermore, we believe that the genetic variants used in our Mendelian randomization analysis serve as an adequate instrument for height for several reasons. One of the assumptions for a valid instrument is that the genetic marker is associated with the exposure phenotype of interest . In our  CORECT) . We alsointerest . Howeverinterest , was notOur study also has several limitations. First, the meta-analyses for the three different cancers are subject to the limitations of the original epidemiologic studies, which could include measurement errors, confounding, selection bias, and random error. Additionally, concerns regarding causal inference from Mendelian randomization studies remain, given the strict criteria required for a valid instrument. In particular, we could not entirely exclude the possible influence of pleiotropic effects on our results since some of the SNPs used in our study might be associated with certain unknown traits that may be related to cancer risk. However, such an influence, if it exists, should be very small, since an instrumental variable constructed using more than 400 height-associated SNPs should be much more strongly associated with adult height than with any other traits. The sample sizes for the analysis of colorectal cancer and for the subgroup analyses of prostate and lung cancers are relatively small, and the relationship between height and these cancers warrants further study using larger sample sizes. Given that most studies showed a linear association between adult height and cancer risk, we assumed a linear association in this analysis, which helped in harmonizing estimates and allowed us to conduct a comprehensive meta-analysis including all relevant studies. A formal test of the shape of the association may be needed in future studies.In summary, this large-scale meta-analysis of prospective studies and Mendelian randomization analysis provide strong evidence for a possible causal association between adult height and the risk of colorectal and lung cancers. Our study suggests that certain biological pathways affecting adult height may also be involved in the etiology of both colorectal and lung cancers.S1 Fig(EPS)Click here for additional data file.S2 Fig(EPS)Click here for additional data file.S3 Fig(EPS)Click here for additional data file.S4 Fig(EPS)Click here for additional data file.S5 Fig(TIF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(DOCX)Click here for additional data file.S4 Table(DOCX)Click here for additional data file.S5 Table(DOCX)Click here for additional data file.S6 Table(DOCX)Click here for additional data file.S7 Table(DOCX)Click here for additional data file.S8 Table(DOCX)Click here for additional data file.S9 Table(DOCX)Click here for additional data file.S1 Text(DOC)Click here for additional data file.