1 INTRODUCTION

Gastric cancer is one of the most prevalent gastrointestinal malignancies and is the third most common cause of cancer-related deaths worldwide.¹ Despite major advances in systemic treatments, surgery remains the only curative option for patients with resectable disease. Multidisciplinary approaches, such as perioperative or neoadjuvant chemotherapies, have been proven to improve both surgical outcomes and long-term survival in patients with high-risk or locally advanced disease.^2-4 However, prolonged preoperative delay imposed by these treatments contributes to risk of progression to unresectable disease or ineligibility for surgery owing to treatment-related toxicities.⁵ In addition, a demand for comprehensive preoperative assessment or intervention has been expanding as the standard of care in recent years.⁶ Such efforts also result in a further delay in curative resection after endoscopic diagnosis. The acceptable interval from endoscopic diagnosis to curative surgery remains largely undetermined and is not routinely described in the international guidelines.

The acceptable preoperative waiting time (PreWT) is an unresolved dilemma because an overshortening of PreWT might be complicated by poor preoperative preparation or conditioning of the patient, while unnecessary lengthening might result in disease progression or psychological stress of the patients.⁷ In the absence of a universal consensus, the impact of PreWT on surgical outcome or prognosis remains to be determined. The United Kingdom Cancer Reform Strategy suggested a standard PreWT of less than 31 days in all cancer patients.⁸ A Korean study by Yun et al. reported that a PreWT of longer than 31 days affected survival in patients with curatively resected gastric cancer in low-volume hospitals.⁹ However, a study from the Netherlands indicated that a longer PreWT of 35–56 days was not associated with a worse survival.¹⁰ In addition to the survival impact, PreWT is recognized as a valid indicator for quality, efficiency, and equity of healthcare, which should be carefully accounted for by healthcare administrations and evaluated in nationwide surveillance.¹¹

Despite several reports about the survival impact of PreWT in patients with resectable gastric cancer, the clinical significance and acceptable interval of PreWT were inconclusive.^12-14 This could be because many of the results were derived from a single institution and may have had cultural and regional disparities, which might not be applicable elsewhere. A large population-based study with an adequate follow-up time is required to address the issue. Furthermore, a detailed elucidation of an exact acceptable value for PreWT may also provide essential information for designing prospective clinical trials and modifying healthcare policies. Therefore, we utilized the population-based cancer registry to investigate the significance of PreWT for patients with resectable stage II–III gastric or gastroesophageal junction (GEJ) cancer and potentiate future interventional clinical trials.

2 MATERIALS AND METHODS

3 RESULTS

3.1 Patients and the distribution of PreWT

A total of 15,473 patients who were diagnosed with gastric or GEJ adenocarcinoma between January 2008 and December 2017 and ≥ D2 gastrectomy were initially screened. From among the evaluable patients (n = 3685), we included patients with a PreWT of 7–118 days in the primary analysis (n = 3059), out of which 1738 patients had cStage II and 1321 patients had cStage III disease. The flow diagram of the study is provided in Figure 1. In the evaluable cohort, the range of PreWT was 1–314 days. In the studied cohort who had a PreWT between 7 and 118 days, the median time was 16 days with an interquartile range (IQR) of 11–24 days. The histogram of the studied population is shown in Figure 2 and that of those with cStage II or III disease in Figure S1.

3.3 PreWT and overall survival

In the studied cohort, patients with a short PreWT of 7–13 days had shorter OS as compared to those with a PreWT of >14 days (median OS by PreWT [days]: 7–13, 2.7 years; 14–20, 3.1 years; 21–27, 3.0 years; 28–34, 4.7 years; 35–31, 3.7 years; 42–48, 3.4 years; 49–118, 2.8 years; p = 0.029) (Figure 3A). Patients with cStage II disease and a PreWT of 7–13 days had a significantly shorter OS as compared to those with a PreWT of >14 days (p = 0.046) (Figure 3B). However, patients with cStage III disease and a PreWT of 49–118 days demonstrated a trend of shorter OS, but the differences were not statistically significant (p = 0.807) (Figure 3C).

3.4 Multivariable Cox and restricted cubic spline regressions

After adjustment for multiple covariates in the Cox regression, PreWT was not statistically significant in patients with cStage II–III disease as an independent prognostic factor. The details of hazard ratios (HRs) and 95% confidence intervals (95% CI) according to different PreWT are shown in Table 2. We observed a reduction in HRs by a PreWT of 35–41 days in all patients (HR = 0.91, 95% CI = 0.66–1.27, p = 0.574) and 35–41 days in cStage III patients (HR = 0.80, 95% CI = 0.49–1.30, p = 0.368) and an increase by a PreWT of 7–13 days in cStage II patients (HR = 1.37, 95% CI = 0.97–1.94, p = 0.075). However, none of these were statistically different as compared to those with a PreWT of 49–118 days.

TABLE 2. Hazard ratios for all-cause mortality.

	All patients (n = 3059)						cStage II (n = 1738)						cStage III (n = 1321)
PreWT, days	n	D	HRa	95% CI	p	p for overallb	n	D	HRa	95% CI	p	p for overallb	n	D	HRa	95% CI	p	p for overallb
7–13	1200	812	1.21	0.94–1.55	0.132	0.719	641	398	1.37	0.97–1.94	0.075	0.501	559	414	0.91	0.64–1.30	0.613	0.352
14–20	858	557	1.16	0.90–1.49	0.244		487	272	1.25	0.88–1.78	0.213		371	285	0.96	0.67–1.38	0.837
21–27	476	301	1.26	0.97–1.64	0.082		279	152	1.32	0.91–1.91	0.138		197	149	1.09	0.75–1.60	0.646
28–34	213	117	0.99	0.73–1.33	0.918		129	58	1.01	0.66–1.53	0.982		84	59	0.89	0.58–1.36	0.580
35–41	124	73	0.91	0.66–1.27	0.574		81	41	0.99	0.63–1.55	0.948		43	32	0.80	0.49–1.30	0.368
42–48	72	44	1.07	0.74–1.57	0.713		49	26	1.14	0.69–1.90	0.615		23	18	0.94	0.53–1.67	0.823
49–118	116	70	1.00				72	36	1.00				44	34	1.00

• Abbreviations: CI, confidence interval; cStage, clinical stage; D, deaths of patients; HR, hazard ratios; n, number of patients; PreWT, preoperative waiting time.
• ^a HRs adjusted for age, sex, treatment site, pathological stage, tumor location, histology grade, size, dissected lymph nodes, resection margin, and receival of adjuvant therapies.
• ^b Overall effect was calculated by multivariable restricted cubic spline regression and compared with Wald test.

In the restricted cubic spline regression under the reference PreWT of 42 days, we observed a reflection point of HRs at day 14 (HR = 1.14, 95% CI = 0.99–1.30), and the HRs gradually decreased with time in all patients (Figure 4A). In cStage II patients, the HRs also decreased with time when the PreWT was prolonged (Figure 4B). To the contrary, the HRs were approximately constant before day 49 and gradually increased with time in cStage III patients (Figure 4C). However, after adjusting for multiple covariates, PreWT was not a statistically significant prognostic factor with time in patients with cStage II–III disease (p = 0.719 in all, p = 0.501 in stage II and p = 0.352 in stage III patients by the Wald test).

4 DISCUSSION

To the best of our knowledge, the present study is one of the largest population-based cohort studies incorporating real-world evidence on the prognostic impact of PreWT in patients with resectable gastric or GEJ cancer. The study indicated that a PreWT of 49–118 days did not independently correlate with a poor prognosis as compared to a PreWT less than 49 days.

Our results agreed with those of previous studies. Furukawa et al. reported that a PreWT up to 90 days did not affect the survival in patients with stage II–III gastric cancer.¹³ Han et al. compared the influence of PreWT in patients with resectable gastric or lung cancer and concluded that the former was unaffected by PreWT in terms of mortality.²⁰ Brenkman et al.'s Netherlands-based study, which included the largest Western patient population in the literature, reported that a PreWT of more than 56 days was not associated with an adverse survival outcome as compared to a PreWT less than 35 days.¹⁰ A similar finding was also reported by Kumazu et al., demonstrating that patients had a comparable OS even when PreWT exceeding 63 days.¹⁴ However, we still observed a trend of shorter OS in patients with a prolonged PreWT, despite the difference not being statistically significant after adjusting for other prognostic factors. The present study confirmed that a prolonged PreWT within an acceptable range was not critical and decisive in terms of prognostic impact in patients with resectable gastric cancer. Nevertheless, our results did not imply that curative surgery could be unconditionally delayed. Weighing the disease risk and therapeutic benefit is still an essential issue when preoperative delay is inevitable in real-world clinical practice.

In light of the success of neoadjuvant or perioperative therapies, the frontline treatment paradigm in gastric cancer has changed in the recent decade.^21-23 In prospective phase III trials, a therapeutic interval of at least 56–63 days was required prior to surgery.^{2-4, 24, 25} Our results revealed that an interval of 49–118 days might be a feasible window period for a comprehensive preoperative assessment and allow time to optimize the condition of the patients without concerns that waiting time might jeopardize survival outcome. In addition, we observed a dynamic change in serial HRs when the PreWT was prolonged by 7 days. The risk of mortality was higher in patients with a PreWT of 7–27 days (adjusted HR = 1.26, 95% CI 0.97–1.64, p = 0.082) and was the least for patients with a PreWT of 35–41 days (adjusted HR = 0.91, 95% CI 0.66–1.27, p = 0.574) as compared to those with a longer PreWT, despite the OS differences not reaching statistical significance. The risk of mortality was also higher in patients with a PreWT less than 30–35 days in the Dutch and Japanese study, consistent with our results.^{10, 13} Indeed, an over-shortening of PreWT might be subjected to result in inadequate preoperative preparation, while a redundant waiting might expose patients to risk of disease progression. Our results echoed a potential optimal preoperative waiting period for clinicians in real-world practices.

The median PreWT was 16 days from endoscopic diagnosis to curative surgery in the current study, which was relatively shorter than that in previous studies reporting a median time ranging from 31 to 45 days.^{9, 10, 13, 14} The possible explanation was the high accessibility to surgery, reliable efficiency, and abundance of medical facilities in Taiwan.^26-28 The UK healthcare administration introduced a waiting time goal of less than 31 days as a quality indicator for excellent clinical service.^{8, 29} Similar appeals were also noted by European, Canadian, and US medical societies for various cancer surgeries.^30-32 Bilimoria et al. reported that 39% of patients with gastric cancer received primary treatment within 30 days from diagnosis, with a median time of 19 days, when the patient was diagnosed and treated at the same hospital in the United States.³³ Another alternative reason was that we excluded patients who were not diagnosed and operated in the same institute to avoid referral time bias; consequently, iatrogenic referral delays were not accounted in the study. In addition to disease concerns, a prolonged waiting time has been proposed to impose a tremendous psychological burden on patients. Song et al. revealed that anxiety while waiting caused complex mental health consequences.³⁴ Although delays in non-emergent cancer surgeries did not impact survival outcomes, they may decrease patient satisfaction and quality of life.^{7, 35, 36}

We observed differences in patient characteristic based on PreWT. Those with a shorter PreWT represented the patient population that was younger, had a more advanced disease, received a deeper level of LN dissection, and was eligible for adjuvant therapies. To the contrary, elderly unfit patients who had a less advanced disease and demanded more preoperative conditioning delayed the surgery and thus contributed to a longer PreWT. Such findings were compatible with those of other retrospective studies.^{9, 10, 12, 14} Furthermore, patients with a shorter PreWT had a higher risk of underestimated disease burden, which was observed as the inconsistency between initial clinical and final pathological staging. The “stage progression” phenomenon highlighted that those with inadequate preoperative assessment had an underestimated disease extent.

The merits of the present study are relatively adequate case numbers and statistical power in a population-based design with long follow-up time. To date, it is the second largest population-based study following the Korean report by Yun et al.⁹ In addition, we used a valid and high-quality cancer registry system to derive analyses beyond a single institute. One limitation was that we excluded those with an extreme PreWT to prevent the influences of outliers. However, we still attempted to conduct sensitivity analyses to confirm the robustness of interpreting results and detailed adjustments on multiple confounding factors for prognostic impacts. Both initial clinical and final pathological staging of the disease were accounted for, and the results reflected direct clinical scenarios encountered by practicing clinicians.

5 CONCLUSION

The present population-based cohort study suggests that a PreWT of 49–118 days does not independently correlate with a poor prognosis as compared to a PreWT less than 49 days in patients with stage II–III gastric or GEJ cancer eligible for curative surgery. The results confirm that a waiting time within an acceptable range is not decisive on prognosis. The study provides a rationale for defining a window period for preoperative neoadjuvant therapies.

AUTHOR CONTRIBUTIONS

Chi-Chieh Yen: Conceptualization (equal); writing – original draft (equal). Yi-Hsin Yang: Conceptualization (equal); formal analysis (equal); writing – original draft (equal). Hsiu-Ying Ku: Data curation (equal); formal analysis (equal). Huang-Ming Hu: Data curation (equal); formal analysis (equal). Su-Shun Lo: Conceptualization (equal); investigation (equal). Hung-Chi Chang: Conceptualization (equal); investigation (equal). Yee Chao: Conceptualization (equal); investigation (equal). Jen-Shi Chen: Conceptualization (equal); investigation (equal). Hsiu-Po Wnag: Conceptualization (equal); investigation (equal). Tsang-En Wang: Conceptualization (equal); investigation (equal). Li-Yuan Bai: Conceptualization (equal); investigation (equal). Ming-Shiang Wu: Conceptualization (equal); investigation (equal). Chia-Jui Yen: Conceptualization (equal); investigation (equal). Li-Tzong Chen: Conceptualization (equal); resources (equal). Yan-Shen Shan: Conceptualization (equal); investigation (equal); resources (equal); writing – review and editing (equal).

FUNDING INFORMATION

This work was funded by the Health Promotion Administration, Ministry of Health and Welfare, Grant no. A1081116. Funded by Tobacco Health and Welfare Taxation. The content of this research may not represent the opinion of the Health Promotion Administration, Ministry of Health and Welfare.

CONFLICT OF INTEREST STATEMENT

The authors have no conflict of interest to declare.

ETHICS STATEMENT

This is a retrospective observational population-based study including patients with standard medical care. All analyzed information underwent a valid de-identification process in TCR and were blinded to all investigators. Therefore, informed consent was waived. The study is approved by the Institutional Review Board (IRB) at National Cheng Kung University Hospital, Tainan, Taiwan with registry serial number A-EX-110-027 in accordance with the Declaration of Helsinki and its amendments.