Short abstract
Objective
To evaluate the association between serum copper levels and lung cancer risk.
Methods
We searched the electronic PubMed, WanFang, CNKI, and SinoMed databases to identify studies including information on serum copper levels and lung cancer. Standard mean differences and corresponding 95% confidence intervals were calculated using Stata 12.0 software. We performed a meta-analysis on the identified studies overall and according to geographic location. We also evaluated heterogeneity among the studies and the occurrence of publication bias.
Results
Thirty-three articles including 3026 cases and 9439 controls were included in our study. The combined results showed that serum copper levels were higher in patients with lung cancer compared with controls without lung cancer, though the results showed high heterogeneity. In a subgroup analysis according to geographic location, significant associations between copper levels and lung cancer were found for both Asian and European populations. No publication bias was detected in this meta-analysis.
Conclusions
High serum copper levels could increase the risk of lung cancer, suggesting that environmental copper exposure may be a risk factor for the development of lung cancer.
Keywords: Serum, copper level, lung cancer, meta-analysis, cancer risk, copper exposure
Introduction
Lung cancer results from the uncontrolled growth of lung tissue cells, which may also cause metastasis.1 Lung cancer is the leading cause of cancer-related death, both in China and worldwide, with 1- and 5-year survival rates of only 42% and 15%, respectively.2 Lung cancer is reportedly the most common cancer among men and women, representing huge social and economic burdens in both developed and developing countries.3 Although antioxidant vitamins and photochemicals have shown protective trends, the roles of trace metals in lung cancer risk remain poorly studied.4,5
Copper is an essential trace metal that plays a key role in maintaining DNA integrity through avoiding oxidative DNA damage or affecting gene mutations.6,7 However, although some studies have reported higher serum copper levels in patients with lung cancer compared with controls,8–10 others found no significant association11,12 or indeed a converse association.13 The effect of serum copper levels on lung cancer risk thus remains controversial. We conducted a meta-analysis to determine the relationship between serum copper levels and lung cancer, and evaluated potential heterogeneities among previous studies.
Methods
Study selection
We performed a comprehensive search of the literature for studies examining serum copper levels and lung cancer up to April 1st, 2018. The PubMed, WanFang, CNKI, and SinoMed databases were searched using the terms “copper concentration” or “copper levels” or “copper” or “Cu” or “trace element” in combination with “lung cancer” or “lung tumor”. Furthermore, references in the relevant articles were also searched to identify other eligible articles.
Inclusion and exclusion criteria
Two investigators (XPZ and QY) independently searched and reviewed articles for eligibility using the following inclusion criteria: 1) studies focusing on patients with lung cancer; 2) observational studies; 3) numbers, mean and standard deviation of serum copper levels for cases and controls available; 4) studies on humans; and 5) studies published in English or Chinese.
Data extraction
Two of the authors (XPZ and QY) independently extracted the following data from the included studies and recorded it in a spreadsheet: 1) first author’s name; 2) publication year; 3) study design; 4) country; 5) number of cases and controls; 6) sex of cases; 7) age; 8) mean and standard deviation of serum copper levels in cases and control; and 9) serum determination method. Other relevant data were also extracted from individual studies.
Statistical analysis
The meta-analysis was carried out using Stata 12.0 software (StataCorp, College Station, TX, USA). Continuous outcomes between serum copper levels and lung cancer were evaluated by calculating the standard mean deviation (SMD) and 95% confidence interval (CI).14 We performed meta-analyses on the identified studies overall and also carried out a subanalysis according to geographic location. The copper concentration in the serum was converted into µmol/L for all studies. Statistical heterogeneity was assessed based on Q and I2 tests.15 The results were combined using a random-effects model. The high between-study heterogeneity was explored by meta-regression analysis.16 Publication bias was evaluated by visual investigation of Begg’s filled funnel plots17 and Egger’s regression asymmetry test.18
Results
Search results and characteristics
The initial screening identified 87, 20, 87, and 79 articles from the PubMed, WanFang, CNKI, and SinoMed databases, respectively. Two additional records were identified through other sources. Figure 1 shows a flow diagram of the study. A total of 33 articles8–13,19–45 involving 3026 lung cancer patients and 9439 controls was finally considered suitable for this study. The characteristics of each study are shown in Table 1.
Table 1.
Study, year [ref] | Country | Age (year) (range or mean ± SD) | Study type |
Lung cancer cases |
Controls |
Method of copper measurement | |||
---|---|---|---|---|---|---|---|---|---|
n | Female (%) | Copper: mean ± SD | n | Copper: mean ± SD | |||||
Sun et al. 1991 [8] | China | 30–75 | Case-control | 91 | 0.00 | 1.267 ± 0.278 (µg/mL) | 138 | 0.921 ± 0.198 (µg/mL) | AAS (IL-951, USA) |
Sun et al. 1991 [8] | China | 30–75 | Case-control | 13 | 100.00 | 1.468 ± 0.416 (µg/mL) | 114 | 1.111 ± 0.324 (µg/mL) | AAS (IL-951, USA) |
Cobanoglu et al. 2010 [13] | Turkey | 54 ± 8.29 | Case-control | 30 | 33.33 | 0.977 ± 0.316 (µg/dL) | 20 | 1.748 ± 0.198 (µg/dL) | UNICAM-929 spectrophotometer (Unicam Ltd., Cambridge, UK) |
Diez et al. 1989 [19] | Spain | 60 ± 7 | Case-control | 64 | 7.81 | 1.4 ± 0.316 (µg/mL) | 100 | 1 ± 0.182 (µg/mL) | AAS (Perkin-Elmer 5.000) |
Huhti et al. 1980 [20] | Finland | 37–80 | Case-control | 149 | 5.37 | 1.42 ± 0.3 (mg/L) | 19 | 1.03 ± 0.26 (mg/L) | AAS (Perkin-Elmer Model 303) |
Jin et al. 2011 [9] | China | 34.9 ± 21.3 | Case-control | 154 | 10.39 | 1.624 ± 0.818 (µg/mL) | 154 | 1.285 ± 0.524 (µg/mL) | AAS (Wako Pure Chemical Industries, Osaka, Japan) |
Oyama et al. 1994 [21] | Japan | 26–83 | Case-control | 109 | 34.86 | 122.9 ± 3.77 (µg/dL) | 53 | 109.5 ± 5.39 (µg/dL) | AAS (Wako Pure Chemical Industries, Osaka, Japan) |
Zowczak et al. 2001 [22] | Poland | 42–87 | Case-control | 14 | 14.29 | 22.9 ± 6.2 (µmol/L) | 18 | 15 ± 1.5 (µmol/L) | Flame AAS (Perkin Elmer) |
Feng et al. 2006 [23] | China | 18–82 | Observation trials | 13 | NA | 19 ± 2.36 (µmol/L) | 36 | 14.92 ± 2.71 (µmol/L) | Flame AAS |
Zhang et al. 1997 [24] | China | 25–80 | Case-control | 64 | 40.63 | 1.512 ± 0.374 (mg/L) | 31 | 1.061 ± 0.157 (mg/L) | AAS |
Jin et al. 2001 [25] | China | 45–70 | Case-control | 40 | 7.50 | 21.7 ± 6.55 (µmol/L) | 46 | 17.2 ± 2.48 (µmol/L) | AAS |
Zhang et al. 1994 [26] | China | 59 ± 9 | Case-control | 40 | 10.00 | 29.67 ± 5.34 (µmol/L) | 24 | 18.84 ± 2.98 (µmol/L) | AAS |
Xu et al. 1993 [11] | China | 56 ± 7.5 | Case-control | 42 | 9.52 | 19.14 ± 4.29 (µmol/L) | 40 | 19.61 ± 1.88 (µmol/L) | AAS |
Zhou et al. 1995 [27] | China | 39–69 | Case-control | 186 | 31.18 | 1.481 ± 0.163 (µg/mL) | 150 | 1.035 ± 0.094 (µg/mL) | AAS |
Chen et al. 1994 [28] | China | 37–72 | Case-control | 58 | 25.86 | 20.1 ± 5.6 (mol/L) | 100 | 18.5 ± 5.1 (mol/L) | AAS (MFX-ID) |
Luo et al. 1996 [29] | China | 40–70 | Case-control | 35 | NA | 17.94 ± 4.09 (µmol/L) | 22 | 9.76 ± 1.89 (µmol/L) | AAS |
Mo et al. 1995 [30] | China | 58.5 | Case-control | 57 | 21.05 | 153.44 ± 33.38 (µg/dL) | 46 | 93.77 ± 12.86 (µg/dL) | AAS |
He et al. 1995 [31] | China | 34–72 | Case-control | 143 | 39.16 | 24.194 ± 9.135 (µmol/L) | 50 | 17.402 ± 5.264 (µmol/L) | AAS |
Wei et al. 2002 [32] | China | 22–76 | Case-control | 79 | 41.77 | 1.093 ± 0.073 (µg/mL) | 32 | 0.867 ± 0.039 (µg/mL) | AAS ( p-100, PE Co., USA) |
Huang et al. 1999 [33] | China | 40–72 | Case-control | 27 | 14.81 | 1.341 ± 0.304 (µg/mL) | 45 | 1.084 ± 0.182 (µg/mL) | AAS |
Zhao et al. 1993 [34] | China | 43–62 | Case-control | 46 | 13.04 | 21.36 ± 4.6 (µmol/L) | 50 | 15.76 ± 4.2 (µmol/L) | AAS (BJKP-36, Beijing, China) |
He et al. 2011 [10] | China | 38–69 | Case-control | 104 | 29.81 | 23.15 ± 3.16 (µmol/L) | 122 | 14.52 ± 1.75 (µmol/L) | AAS |
Chen et al. 1998 [35] | China | 47–72 | Case-control | 43 | 32.56 | 19.08 ± 3.41 (µmol/L) | 180 | 13.85 ± 2.36 (µmol/L) | AAS (A670, Shimadzu, Japan) |
Liang et al. 1992 [36] | China | 61 | Case-control | 57 | 21.05 | 28.75 ± 9.7 (µmol/L) | 80 | 19.76 ± 3.56 (µmol/L) | AAS (WFX-ID, China) |
Huang et al. 1998 [12] | China | 25–65 | Case-control | 136 | 19.12 | 21.453 ± 5.783 (µmol/L) | 7101 | 20.713 ± 5.508 (µmol/L) | AAS (AA670/C2H2, Shimadzu) |
Wang et al. 2003 [37] | China | 28–69 | Case-control | 50 | 40.00 | 1.04 ± 0.2 (µg/L) | 60 | 0.77 ± 0.22 (µg/L) | AAS |
Cheng et al. 2011 [38] | China | 37–68 | Case-control | 197 | 32.99 | 1.19 ± 0.13 (µmol/L) | 93 | 0.87 ± 0.35 (µmol/L) | AAS |
Xie et al. 2000 [39] | China | 35–68 | Case-control | 64 | 45.31 | 25.3 ± 6.3 (µmol/L) | 100 | 22.1 ± 1.7 (µmol/L) | AAS |
Du et al. 1996 [40] | China | 22–73 | Case-control | 73 | 31.51 | 21.3 ± 4.3 (µmol/L) | 63 | 15.3 ± 3.4 (µmol/L) | AAS |
Zhu et al. 1997 [41] | China | NA | Case-control | 56 | NA | 21.05 ± 3.56 (µmol/L) | 118 | 16.01 ± 2.13 (µmol/L) | AAS (3030, Perkin Elmer Zeeman, USA) |
Zhang et al. 2000 [42] | China | 25–77 | Case-control | 310 | 17.74 | 1.151 ± 0.264 (µg/mL) | 48 | 1.068 ± 0.233 (µg/mL) | AAS (180-80, Shimadzu, Japan) |
Hu et al. 2000 [43] | China | 36–77 | Case-control | 56 | 17.86 | 1.508 ± 0.379 (µg/mL) | 60 | 1.403 ± 0.148 (µg/mL) | AAS |
Guo et al. 1994 [44] | China | 55.1 | Case-control | 26 | 26.92 | 2.81 ± 1.54 (µg/mL) | 26 | 0.82 ± 0.21 (µg/mL) | AAS (AA-40p, Varian, USA) |
Han et al. 1999 [45] | China | NA | Case-control | 400 | NA | 1.12 ± 0.43 (µg/mL) | 100 | 0.87 ± 0.26 (µg/mL) | AAS (3030, PE Co., USA) |
AAS, atomic absorption spectrophotometry; SD, standard deviation; NA, not available
Serum copper levels and risk of lung cancer
In the overall analysis, lung cancer patients had significantly higher serum copper levels than controls (summary SMD=1.103, 95%CI=1.040–1.165, Z=34.55, P for Z test <0.001), with significant between-study heterogeneity (I2=96.4%, P<0.001) (Figure 2).
Thirty-two of the included 33 articles were case-control studies, and the result for these was consistent with the overall result (summary SMD=1.099, 95%CI=1.036–1.162, Z=34.30, P for Z test <0.001). In a stratified analysis according to geographic location, the associations between serum copper levels and lung cancer were significant for both Asian (summary SMD=1.078, 95%CI=1.013–1.142, Z=32.88, P for Z test <0.001] and European populations (summary SMD=1.568, 95%CI=1.292–1.845, Z=11.13, P for Z test <0.001). Detailed results are shown in Table 2.
Table 2.
Study | No. of studies | SMD (95% CI) |
Z test |
Heterogeneity test |
||
---|---|---|---|---|---|---|
Z | P-value | I2 (%) | P-value | |||
All | 33 | 1.103 (1.040–1.165) | 34.55 | <0.001 | 96.4 | <0.001 |
Geographic location | ||||||
Europe | 3 | 1.568 (1.292–1.845) | 11.13 | <0.001 | 0.0 | 0.444 |
Asia | 30 | 1.078 (1.013–1.142) | 32.88 | <0.001 | 96.6 | <0.001 |
Study type | ||||||
Case-control | 32 | 1.099 (1.036–1.162) | 34.30 | <0.001 | 96.5 | <0.001 |
Observation trials | 1 | – | – | – | – | – |
SMD, standard mean deviation; CI, confidence interval
Between-study heterogeneity
Significant evidence of between-study heterogeneity was detected when we pooled the overall results. We therefore performed univariate meta-regression analysis to explore the source of the high heterogeneity. No specific covariate (publication year, geographic location, case number) accounted for this high heterogeneity.
Publication bias and sensitivity analysis
Egger’s regression asymmetry test (P=0.103) and Begg’s filled funnel plots (Figure 3) detected no publication bias.
Sensitivity analysis showed no apparent effect on the overall merged SMD after deleting any individual study, indicating that no single study influenced the overall effect (Figure 4).
Discussion
Previous analyses have shown inconsistent results regarding the relationship between serum copper levels and lung cancer, probably due to limited sample sizes. We therefore conducted a meta-analysis of pooled data to obtain a comprehensive result and showed that elevated serum copper levels may increase the risk of lung cancer. Furthermore, serum copper levels were higher in lung cancer patients than in controls in both European and Asian populations.
A previous meta-analysis suggested that patients with thyroid cancer had higher copper levels than healthy controls.46 Another meta-analysis showed that serum copper levels were markedly higher in patients with bladder cancer compared with individuals without bladder cancer.47 Furthermore, a recent study found higher serum copper levels in patients with cervical cancer than in controls.48 The current results are consistent with the above studies. The reason why serum copper levels may be elevated in patients with lung cancer may be related to copper metabolism. Serum copper levels in healthy people are associated with ceruloplasmin,49 which is normally catabolized in the liver following cleavage of its terminal sialic acid chains by neuraminidase.50 It has been suggested51 that ceruloplasmin may be resialylated at the tumor cell surface or in the peripheral blood in patients with neoplasms, thus inhibiting its catabolism and potentially explaining the increase in serum copper levels in patients with malignant tumors.
This meta-analysis had several important strengths. First, the study included a large numbers of cases and participants, yielding a comprehensive result. Second, removing each individual study from the analysis had no apparent effect on the overall merged SMD, indicating that the results were stable. Third, no small study effect was detected by Egger’s regression asymmetry test or Begg’s filled funnel plots.
However, several limitations also need to be considered when interpreting the results. First, most of the included studies involved Asian populations and only three studies were from Europe. Although subgroup analysis identified significant associations between serum copper and lung cancer in both these subgroups, future studies in European and other populations are warranted to clarify the relationship between serum copper levels and lung cancer risk. Second, lung cancer is a complex disease with a variety of etiologic factors, including environmental and genetic factors. It is therefore possible that other factors may have influenced the results. Third, although most of the included studies measured copper levels using atomic absorption spectrophotometry, the use of instruments produced by different companies could have led to inconsistent measurements. Finally, significant heterogeneity between studies was observed in this meta-analysis. However, the heterogeneity was mainly related to the strength of the association rather than the direction of the risk estimate, suggesting that the findings in relation to the investigated outcome were promising. Furthermore, an investigation of potential covariates by meta-regression analysis found no significant contribution of publication year, geographic location, sex, or case number to the high between-study heterogeneity. No single study accounted for the significant between-study heterogeneity or influenced the overall result according to sensitivity analysis.
Conclusions
This meta-analysis concluded that serum copper levels tend to be higher in patients with lung cancer than in controls without lung cancer. Environmental copper exposure may thus increase the risk of lung cancer.
Declaration of conflicting interest
The authors declare that there is no conflict of interest.
Funding
This work was supported by the “Psychological therapy and related research on chronic pain in cancer patients” (C2015-86).
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