Abstract
The complex relationship between measured leukocyte telomere length (LTL), genetically predicted LTL (gTL), and carcinogenesis is exemplified by lung cancer. We previously reported associations between longer pre-diagnostic LTL, gTL, and increased lung cancer risk among European and East Asian populations. However, we had limited statistical power to examine the associations among never smokers by gender and histology.
To investigate further, we conducted nested case–control analyses on an expanded sample of never smokers from the prospective Shanghai Women's Health Studies (798 cases and 792 controls) and Shanghai Men's Health Studies (161 cases and 162 controls). We broke the case–control matching and used multivariable unconditional logistic regression models to estimate the ORs and 95% confidence intervals (CI) of incident lung cancer and adenocarcinoma (LUAD), in relation to LTL measured using quantitative PCR and gTL determined using a polygenic score. In addition, we conducted Mendelian randomization (MR) using MR-PRESSO.
We found striking dose–response relationships between longer LTL and gTL, and increased lung cancer risk among never-smoking women (P trendLTL = 4×10−6; P trendgTL = 3×10−4). Similarly, among never-smoking men, longer measured LTL was associated with over triple the risk compared with those with the shortest (OR, 3.48; 95% CI, 1.85–6.57). The overall results were similar for LUAD among women and men. MR analyses supported causal associations with LUAD among women (OR1 SD gTL, 1.19; 95% CI, 1.03–1.37; P = 0.03).
Longer pre-diagnostic LTL is associated with increased lung cancer risk among never smokers.
Our findings firmly support the role of longer telomeres in lung carcinogenesis.
Introduction
Telomeres are repetitive (TTAGGG)n sequences at the ends of linear chromosomes that protect the genome against progressive decay. Telomere length measured in circulating leukocytes [leukocyte telomere length (LTL)] is an established biological marker of genomic instability that captures both the endogenous factors related to biological aging and the cumulative burden of exogenous exposures.
The complex relationship between measured LTL, genetically predicted LTL (gTL), and carcinogenesis is exemplified by lung cancer (1–4). In a pooled analysis of three prospective studies (1), we previously reported consistent associations between longer pre-diagnostic LTL and gTL and increased lung cancer risk among European and East Asian populations (1, 5, 6), which was independently supported by other prospective studies (3, 4). These positive dose–response relationships defy the traditional expectation that shorter telomeres are detrimental in relation to early cancer development. As such, further replication in larger, diverse study populations is needed.
Although we previously found a positive LTL-lung cancer association among never-smoking Chinese women living in an urban setting (1), the limited sample size and lack of LTL data among comparable men limited our ability to investigate the relationships among subgroups defined by gender, smoking status, and histologic subtype. To investigate further, we assessed the relationship between pre-diagnostic LTL, gTL, and lung cancer in a new, expanded sample of never-smoking Chinese women and men from the prospective Shanghai Women's Health Studies (SWHS) and Shanghai Men's Health Studies (SMHS). We conducted additional Mendelian randomization (MR) and gTL analyses to further support the findings.
Materials and Methods
Details of the source population, prospective cohort study design, case ascertainment, and control selection were previously described (1, 7, 8). We conducted nested case–control analyses separately for never-smoking Chinese women from the SWHS (798 incident lung cancer cases and 792 cancer-free controls) and never-smoking men from the SMHS (161 incident lung cancer cases and 162 cancer-free controls). The controls were matched to cases on age (±2 years) and date of blood sample collection (±3 months). Compared with the previously published analyses of 215 cases and 215 controls (1), our current expanded SWHS sample included 583 more lung cancer cases (271% increase) and 577 more controls (268% increase). The SWHS and SMHS were sampled from the same source population and urban geographic region in the Shanghai metropolitan area. Written informed consent was obtained from all subjects prior to their participation.
To maximize statistical power, we broke the case–control matching and used unconditional logistic regression models to estimate the ORs and 95% confidence intervals (CI) of incident lung cancer and adenocarcinoma (LUAD) separately, in relation to quartiles of LTL measured using quantitative PCR (9). We adjusted for matching factors and confounders including age at recruitment, body mass index (BMI), education, alcohol intake, and exposure to secondhand smoke (women only). Among replicate quality control samples, the qPCR assay had an intraclass correlation of 95.6% and coefficient of variation of 7.7%.
In parallel, we generated a polygenic score (PGS) for longer gTL (2) among SWHS only, using 32 previously identified genetic variants associated with LTL in a multiethnic population from TOPMed (10). The PGS was then categorized into quartiles based on the distribution of the controls and analyzed in relation to lung cancer and LUAD risk, adjusted for age and 10 principal components to account for population stratification.
To assess potential causal associations between gTL and lung cancer, we conducted MR analyses using the same 32 genetic variants used to derive the PGS (10). Here, we used MR-PRESSO (11), a robust approach that estimates causal effects after removing variants detected with evidence of pleiotropic effects. P values < 0.05 were considered statistically significant.
Human subjects Institutional Review Board
The protocol for this study was approved by the NCI Institutional Review Board (IRB). The SWHS and SMHS were approved by the IRBs of Vanderbilt University and the Shanghai Cancer Institute in accordance with the Declaration of Helsinki.
Data availability
The data that support the findings of this study are available on request from the senior author. The data are not publicly available due to privacy or ethical restrictions.
Results
The characteristics of the SWHS and SMHS study populations are shown in Table 1. The average age at recruitment was 57.85 years for women and 62.50 years for men. There was no evidence for case–control differences for age or any other variable among women and men.
. | I) Women (798 lung cancer cases; 792 controls) . | . | II) Men (161 lung cancer cases; 162 controls) . | . | ||||||
---|---|---|---|---|---|---|---|---|---|---|
. | Cases . | Controls . | P value . | Cases . | Controls . | P value . | ||||
Age at recruitment, years, mean (SD) | 56.8 | (8.9) | 56.9 | (8.9) | 0.70 | 62.3 | (8.4) | 62.7 | (8.5) | 0.60 |
Secondhand smoke exposure at home or outside, n, % | ||||||||||
Unexposed (<1 hour/week) | 125 | (15.7) | 111 | (14.0) | 0.40 | No data | ||||
Exposed (≥1 hour/week) | 622 | (77.9) | 638 | (80.6) | No data | |||||
Missing | 51 | (6.4) | 43 | (5.4) | No data | |||||
BMI, kg/m2, overall mean (SD) and categorical n, % | 24.3 | (3.5) | 24.6 | (3.4) | 0.04 | 23.7 | (2.6) | 23.5 | (3.0) | 0.43 |
<18.5 | 17 | (2.1) | 18 | (2.3) | 0.40 | 5 | (3.1) | 10 | (6.2) | 0.57 |
18.5–<25 | 481 | (60.3) | 444 | (56.1) | 111 | (68.9) | 106 | (65.4) | ||
25–<30 | 255 | (32.0) | 286 | (36.1) | 37 | (23.0) | 41 | (25.3) | ||
30–35 | 40 | (5.0) | 37 | (4.7) | 4 | (2.5) | 5 | (3.1) | ||
35≤ | 5 | (0.6) | 7 | (0.9) | 0 | (0.0) | 0 | (0.0) | ||
Missing | 0 | 0.0 | 0 | (0.0) | 4 | (2.5) | 0 | (0.0) | ||
Education, n, % | ||||||||||
College or more | 124 | (15.5) | 121 | (15.3) | 0.90 | 70 | (43.5) | 67 | (41.4) | 0.30 |
High school | 181 | (22.7) | 171 | (21.6) | 36 | (22.4) | 44 | (27.2) | ||
Middle school | 235 | (29.4) | 231 | (29.2) | 36 | (22.4) | 38 | (23.5) | ||
Elementary or less | 258 | (32.3) | 269 | (34.0) | 15 | (9.3) | 13 | (8.0) | ||
Missing | 0 | (0.0) | 0 | (0.0) | 4 | (2.5) | 0 | (0.0) | ||
Regular alcohol consumption, n, % | ||||||||||
Yes | 14 | (1.8) | 19 | (2.4) | 0.40 | 24 | (15.3) | 30 | (18.5) | 0.10 |
No | 784 | (98.2) | 773 | (97.6) | 133 | (84.7) | 132 | (81.5) |
. | I) Women (798 lung cancer cases; 792 controls) . | . | II) Men (161 lung cancer cases; 162 controls) . | . | ||||||
---|---|---|---|---|---|---|---|---|---|---|
. | Cases . | Controls . | P value . | Cases . | Controls . | P value . | ||||
Age at recruitment, years, mean (SD) | 56.8 | (8.9) | 56.9 | (8.9) | 0.70 | 62.3 | (8.4) | 62.7 | (8.5) | 0.60 |
Secondhand smoke exposure at home or outside, n, % | ||||||||||
Unexposed (<1 hour/week) | 125 | (15.7) | 111 | (14.0) | 0.40 | No data | ||||
Exposed (≥1 hour/week) | 622 | (77.9) | 638 | (80.6) | No data | |||||
Missing | 51 | (6.4) | 43 | (5.4) | No data | |||||
BMI, kg/m2, overall mean (SD) and categorical n, % | 24.3 | (3.5) | 24.6 | (3.4) | 0.04 | 23.7 | (2.6) | 23.5 | (3.0) | 0.43 |
<18.5 | 17 | (2.1) | 18 | (2.3) | 0.40 | 5 | (3.1) | 10 | (6.2) | 0.57 |
18.5–<25 | 481 | (60.3) | 444 | (56.1) | 111 | (68.9) | 106 | (65.4) | ||
25–<30 | 255 | (32.0) | 286 | (36.1) | 37 | (23.0) | 41 | (25.3) | ||
30–35 | 40 | (5.0) | 37 | (4.7) | 4 | (2.5) | 5 | (3.1) | ||
35≤ | 5 | (0.6) | 7 | (0.9) | 0 | (0.0) | 0 | (0.0) | ||
Missing | 0 | 0.0 | 0 | (0.0) | 4 | (2.5) | 0 | (0.0) | ||
Education, n, % | ||||||||||
College or more | 124 | (15.5) | 121 | (15.3) | 0.90 | 70 | (43.5) | 67 | (41.4) | 0.30 |
High school | 181 | (22.7) | 171 | (21.6) | 36 | (22.4) | 44 | (27.2) | ||
Middle school | 235 | (29.4) | 231 | (29.2) | 36 | (22.4) | 38 | (23.5) | ||
Elementary or less | 258 | (32.3) | 269 | (34.0) | 15 | (9.3) | 13 | (8.0) | ||
Missing | 0 | (0.0) | 0 | (0.0) | 4 | (2.5) | 0 | (0.0) | ||
Regular alcohol consumption, n, % | ||||||||||
Yes | 14 | (1.8) | 19 | (2.4) | 0.40 | 24 | (15.3) | 30 | (18.5) | 0.10 |
No | 784 | (98.2) | 773 | (97.6) | 133 | (84.7) | 132 | (81.5) |
Note: P values < 0.05 were considered statistically significant.
We found strong associations between longer pre-diagnostic LTL, gTL, and increased lung cancer risk among never-smoking women and men from urban Shanghai, China. Among never-smoking women, we found a statistically significant dose–response relationship between longer measured LTL and increased risk of lung cancer (Ptrend = 4×10−6). The highest quartile of LTL was associated with over double the risk compared with the lowest quartile (OR, 2.10; 95% CI, 1.52–2.90; P = 6×10−6; Table 2). The findings among women were similar when examining LUAD (Ptrend = 1×10−3), the lung cancer subtype most commonly found among never smokers (Table 2). Similarly, we found a dose–response relationship between longer gTL and increased lung cancer (Ptrend = 3×10−4) and LUAD risk (Ptrend = 4×10−4; Table 3). The LUAD findings were further supported by MR analyses (ORcausal, per 1 SD gTL, 1.19; 95% CI, 1.03–1.37; P = 0.03). However, the MR results of overall lung cancer were in the same direction but non-significant (ORcausal, per 1 SD gTL, 1.09; 95% CI, 0.96–1.22; P = 0.14).
. | . | A) Women (798 cases; 792 controls) . | B) Men (161 cases; 162 controls) . | ||||||||||
---|---|---|---|---|---|---|---|---|---|---|---|---|---|
Quartile . | Quartiles, LTL cutoffs . | No. of cases . | No. of controls . | OR . | 95% CI Lower . | 95% CI Upper . | P value/Ptrend . | No. of cases . | No. of controls . | OR . | 95% CI lower . | 95% CI upper . | P value/Ptrend . |
I) Overall lung cancer | |||||||||||||
1 | <0.487 | 110 | 174 | 1.00 | 39 | 63 | 1.00 | ||||||
2 | 0.487–<0.575 | 177 | 203 | 1.45 | 1.05 | 2.02 | 0.03 | 30 | 35 | 1.37 | 0.71 | 2.62 | 0.35 |
3 | 0.575–<0.724 | 242 | 213 | 1.76 | 1.28 | 2.43 | 5×10−4 | 53 | 27 | 3.48 | 1.85 | 6.57 | 1×10−4 |
4 | ≥0.724 | 269 | 202 | 2.10 | 1.52 | 2.90 | 6×10−6 | 39 | 37 | 1.85 | 0.99 | 3.44 | 0.05 |
Ordinal trend | 1.26 | 1.14 | 1.39 | 4×10−6 | 1.33 | 1.09 | 1.62 | 6×10−3 | |||||
II) Lung adenocarcinoma | |||||||||||||
1 | <0.487 | 52 | 174 | 1.00 | 32 | 63 | 1.00 | ||||||
2 | 0.487–<0.575 | 105 | 203 | 1.61 | 1.06 | 2.43 | 0.02 | 28 | 35 | 1.53 | 0.78 | 3.03 | 0.22 |
3 | 0.575–<0.724 | 140 | 213 | 1.88 | 1.25 | 2.81 | 2×10−3 | 48 | 27 | 4.02 | 2.06 | 7.84 | 4×10−5 |
4 | ≥0.724 | 150 | 202 | 2.00 | 1.34 | 3.00 | 8×10−4 | 27 | 37 | 1.54 | 0.78 | 3.03 | 0.22 |
Ordinal trend | 1.22 | 1.08 | 1.37 | 1×10−3 | 1.28 | 1.04 | 1.58 | 0.02 |
. | . | A) Women (798 cases; 792 controls) . | B) Men (161 cases; 162 controls) . | ||||||||||
---|---|---|---|---|---|---|---|---|---|---|---|---|---|
Quartile . | Quartiles, LTL cutoffs . | No. of cases . | No. of controls . | OR . | 95% CI Lower . | 95% CI Upper . | P value/Ptrend . | No. of cases . | No. of controls . | OR . | 95% CI lower . | 95% CI upper . | P value/Ptrend . |
I) Overall lung cancer | |||||||||||||
1 | <0.487 | 110 | 174 | 1.00 | 39 | 63 | 1.00 | ||||||
2 | 0.487–<0.575 | 177 | 203 | 1.45 | 1.05 | 2.02 | 0.03 | 30 | 35 | 1.37 | 0.71 | 2.62 | 0.35 |
3 | 0.575–<0.724 | 242 | 213 | 1.76 | 1.28 | 2.43 | 5×10−4 | 53 | 27 | 3.48 | 1.85 | 6.57 | 1×10−4 |
4 | ≥0.724 | 269 | 202 | 2.10 | 1.52 | 2.90 | 6×10−6 | 39 | 37 | 1.85 | 0.99 | 3.44 | 0.05 |
Ordinal trend | 1.26 | 1.14 | 1.39 | 4×10−6 | 1.33 | 1.09 | 1.62 | 6×10−3 | |||||
II) Lung adenocarcinoma | |||||||||||||
1 | <0.487 | 52 | 174 | 1.00 | 32 | 63 | 1.00 | ||||||
2 | 0.487–<0.575 | 105 | 203 | 1.61 | 1.06 | 2.43 | 0.02 | 28 | 35 | 1.53 | 0.78 | 3.03 | 0.22 |
3 | 0.575–<0.724 | 140 | 213 | 1.88 | 1.25 | 2.81 | 2×10−3 | 48 | 27 | 4.02 | 2.06 | 7.84 | 4×10−5 |
4 | ≥0.724 | 150 | 202 | 2.00 | 1.34 | 3.00 | 8×10−4 | 27 | 37 | 1.54 | 0.78 | 3.03 | 0.22 |
Ordinal trend | 1.22 | 1.08 | 1.37 | 1×10−3 | 1.28 | 1.04 | 1.58 | 0.02 |
Note: Multivariable unconditional logistic regression models were used to estimate the ORs and 95% CIs of incident overall lung cancer and LUAD in relation to quartiles of measured LTL, adjusted for age at recruitment (continuous), BMI (continuous), education, alcohol intake (no vs. yes); exposure to secondhand smoke (ever vs. never, women only). Ptrends were estimated using continuous adjusted T/S ratio and ordinal quartiles. P values < 0.05 were considered statistically significant.
. | I) Overall lung cancer . | II) Lung adenocarcinoma . | ||
---|---|---|---|---|
. | (735 cases; 737 controls) . | (415 cases; 737 controls) . | ||
Quartiles, PGS for longer gTL . | OR (95% CI) . | P value/Ptrend . | OR (95% CI) . | P value/Ptrend . |
1 | Ref | Ref | ||
2 | 1.11 (0.82–1.51) | 0.50 | 1.23 (0.84–1.80) | 0.28 |
3 | 1.23 (0.91–1.67) | 0.17 | 1.36 (0.94–1.98) | 0.10 |
4 | 1.63 (1.21–2.19) | 1×10−3* | 1.90 (1.33–2.72) | 5×10−4* |
Continuousa | 1.20 (1.09–1.33) | 3×10−4* | 1.25 (1.11–1.41) | 4×10−4* |
. | I) Overall lung cancer . | II) Lung adenocarcinoma . | ||
---|---|---|---|---|
. | (735 cases; 737 controls) . | (415 cases; 737 controls) . | ||
Quartiles, PGS for longer gTL . | OR (95% CI) . | P value/Ptrend . | OR (95% CI) . | P value/Ptrend . |
1 | Ref | Ref | ||
2 | 1.11 (0.82–1.51) | 0.50 | 1.23 (0.84–1.80) | 0.28 |
3 | 1.23 (0.91–1.67) | 0.17 | 1.36 (0.94–1.98) | 0.10 |
4 | 1.63 (1.21–2.19) | 1×10−3* | 1.90 (1.33–2.72) | 5×10−4* |
Continuousa | 1.20 (1.09–1.33) | 3×10−4* | 1.25 (1.11–1.41) | 4×10−4* |
Note: Multivariable unconditional logistic regression models were used to estimate the ORs and 95% CIs of incident overall lung cancer and LUAD in relation to quartiles of gTL (PGS), adjusted for age at recruitment (continuous), and 10 principal components. *P values < 0.05 were considered statistically significant.
aper 1 standard deviation increase in continuous gTL.
PGS was standardized and categorized into quartiles based on the distribution among controls.
Among never-smoking Chinese men, the association between LTL and lung cancer risk was not monotonic (Table 2). Here, we found that the third quartile of LTL was associated with over triple the risk of lung cancer compared with the lowest quartile (OR, 3.48; 95% CI, 1.85–6.57; P = 1×10−4), which was similarly strong for LUAD (OR, 4.02; 95% CI, 2.06–7.84; P = 4×10−5). We did not have genetic data to conduct PGS or MR analyses among men.
Discussion
We conducted a prospective study of both phenotypically measured LTL and gTL, and lung cancer risk among never smokers who lived in Shanghai, a metropolis with similar types and patterns of vehicular and industrial air pollution as other urban megacities in Asia. Our findings were in agreement with previous prospective cohort and genome-wide association studies (GWAS) that reported positive dose–response relationships between LTL, gTL, and lung cancer risk among European and East Asian populations (1–6, 12). The intriguing positive associations are apparently consistent and robust across the literature among diverse subgroups defined by geographic region, sex, race/ethnicity, and smoking status.
The positive association between longer telomeres and increased lung cancer risk defies the traditional expectations of telomere dynamics in cancer development. Longer telomeres are generally reflective of decreased genomic instability, biological aging, and exposure to environmental stressors. To explain this phenomenon, we hypothesized that longer LTL may reflect or lead to delayed senescence or greater replicative potential of precancerous cells, which allows them to accumulate somatic genomic abnormalities that potentially drive carcinogenesis and cancer progression (5). Further, a seminal paper posited that a series of LTL-independent mutational hits at the stem-cell level may generate a clone with replicative advantage and subsequently, additional LTL-related mutational hits potentially transforms the expanding clone into cancer (13). A recent investigation may yield some clues into the paradoxical telomere–cancer relationship. A family study of 17 Human Protection of Telomeres 1 (POT1) mutation carriers and 21 noncarrier relatives found that POT1 mutations associated with longer telomere length predisposed carriers to a familial clonal hematopoiesis syndrome that was associated with various benign and malignant solid neoplasms (14). Nevertheless, the biological mechanism underlying this association is unclear and warrants further investigation.
Our study had notable strengths. First, we conducted one of the limited number of prospective studies with measured LTL that focused on never smokers, a group in which the burden of lung cancer is understudied but is becoming increasingly appreciated. Recently, the United Kingdom National Cancer Research Institute Lung Study Group presented recommendations for a national research strategy aimed at improving outcomes for lung cancer patients who never smoked (15). Second, the current study included an expanded sample of female and male lung cancer cases compared with the previous pooled study (1), which allowed for the analysis of the LUAD subtype among never smokers, thus providing deeper insight into lung carcinogenesis. Third, our prospective study analyzed LTL measured in pre-diagnostic biospecimens. Therefore, our findings were robust against disease-bias (reverse causation) that is inherent in many retrospective studies of biomarkers in which biospecimens were collected after disease diagnosis and treatment.
Our study had some limitations. Although, we had an expanded study sample, we did not have sufficient case numbers to analyze squamous cell carcinoma (SCC). However, SCC is more common among smokers and previous studies of telomere length and lung cancer found limited evidence for associations with SCC. Furthermore, we did not have information on leukocyte differentials. However, our findings were consistent with previous studies of telomere length and lung cancer risk, which increased our confidence in our results.
In summary, we found robust prospective associations between longer pre-diagnostic and gTL and increased lung cancer risk. Our findings validate the results from previous prospective cohort studies and GWAS and expand further by examining the associations with the LUAD subtype among never-smoking men and women. Our findings firmly support the role of longer telomeres in the pathogenesis of lung cancer. However, the precise biological mechanism underlying these paradoxical findings requires further integration of experimental and population studies.
Authors' Disclosures
X.O. Shu reports grants from NCI during the conduct of the study. R. Cawthon reports other support from NCI during the conduct of the study; in addition, R. Cawthon has a patent 8048631 issued to University of Utah. G. Yang reports grants from NIH during the conduct of the study. No disclosures were reported by the other authors.
Authors' Contributions
J.Y.Y. Wong: Conceptualization, methodology, writing–original draft. X.-O. Shu: Resources, methodology, writing–review and editing. W. Hu: Data curation, formal analysis, methodology, writing–review and editing. B. Blechter: Formal analysis, writing–original draft. J. Shi: Formal analysis, methodology. K. Wang: Formal analysis. R. Cawthon: Methodology. Q. Cai: Methodology, writing–review and editing. G. Yang: Methodology, writing–review and editing. M.L. Rahman: Writing–review and editing. B.-t. Ji: Data curation, writing–review and editing. Y. Gao: Writing–review and editing. W. Zheng: Resources, methodology, writing–review and editing. N. Rothman: Conceptualization, resources, supervision, writing–review and editing. Q. Lan: Conceptualization, resources, supervision, writing–review and editing.
Acknowledgments
The current study was supported by the intramural research program of the NCI, NIH (Principal investigator: Q. Lan, Project Number: 1ZIACP010120–27).
The SWHS cohort was funded by NIH research grant R01 CA70867 (to W. Zheng) and by Intramural Research Program contract N02 CP1101066.
The SMHS cohort was funded by the NCI, NIH [grant numbers R01 CA082729, UM1 CA173640 (to X.O. Shu)], and partially by the State Key Project Specialized for Infectious Diseases, China [grant numbers 2008ZX10002–015, 2012ZX10002008–002].
We thank all participants of the SWHS and SMHS.
The publication costs of this article were defrayed in part by the payment of publication fees. Therefore, and solely to indicate this fact, this article is hereby marked “advertisement” in accordance with 18 USC section 1734.